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The science of systems that learn from data and experience. Its scope runs from the statistical and mathematical foundations of learning, including generalization, expressivity, and computational limits, through the design of learning algorithms, deep learning, reinforcement learning, and probabilistic methods, to the empirical study of large models and the trustworthiness, interpretability, and societal impact of learned systems.

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Convex OptimizationLinear algebraOptimization·Captain: mikedeng1

Adaptive Subgradient Methods for Online Learning and Stochastic Optimization 2: Full-Matrix AdaGrad's Regret Is Bounded by tr(G_T^{1/2})Research Paper

Motivation

Online convex optimization is the standard model for learning from a stream of data: in round ttt a learner commits to a point xtx_txt​ in a convex set X⊆Rd\mathcal X\subseteq\mathbb R^dX⊆Rd, a convex loss ftf_tft​ is revealed, and the learner pays ft(xt)+φ(xt)f_t(x_t)+\varphi(x_t)ft​(xt​)+φ(xt​), where φ\varphiφ is a fixed regularizer. Subgradient methods for this model use one fixed geometry (usually the Euclidean one) for every coordinate and every direction, regardless of the data. Duchi, Hazan and Singer (JMLR 12 (2011) 2121–2159) introduced ADAGRAD, which adapts the geometry to the observed subgradients. The diagonal version is one of the most widely used optimizers in machine learning and the ancestor of RMSProp and Adam. This mission formalizes the paper's full-matrix version: the proximal term is built from the matrix square root of the accumulated outer products of the subgradients, so the method adapts to correlated directions, not only to individual coordinates.

The analysis extends the regret bounds of primal-dual subgradient methods and regularized dual averaging (Nesterov 2009; Xiao 2010) and of composite mirror descent to proximal functions that change over time, and combines them with trace inequalities for the matrix square root. Concurrent work by McMahan and Streeter (COLT 2010) studied closely related adaptive proximal methods.

Setting

Vectors live in Rd\mathbb R^dRd with the Euclidean inner product ⟨⋅,⋅⟩\langle\cdot,\cdot\rangle⟨⋅,⋅⟩ and norm ∥⋅∥2\|\cdot\|_2∥⋅∥2​. In round t=1,2,…t=1,2,\dotst=1,2,… the learner plays xt∈Xx_t\in\mathcal Xxt​∈X and observes a subgradient gt∈∂ft(xt)g_t\in\partial f_t(x_t)gt​∈∂ft​(xt​), that is, ft(y)≥ft(xt)+⟨gt,y−xt⟩f_t(y)\ge f_t(x_t)+\langle g_t,y-x_t\rangleft​(y)≥ft​(xt​)+⟨gt​,y−xt​⟩ for all yyy. The regret against a comparator x∗∈Xx^*\in\mathcal Xx∗∈X is

Rϕ(T)=∑t=1T[ft(xt)+φ(xt)−ft(x∗)−φ(x∗)].R_\phi(T)=\sum_{t=1}^T\big[f_t(x_t)+\varphi(x_t)-f_t(x^*)-\varphi(x^*)\big].Rϕ​(T)=t=1∑T​[ft​(xt​)+φ(xt​)−ft​(x∗)−φ(x∗)].

The outer product matrix is Gt=∑τ=1tgτgτ⊤G_t=\sum_{\tau=1}^t g_\tau g_\tau^\topGt​=∑τ=1t​gτ​gτ⊤​, a symmetric positive semidefinite d×dd\times dd×d matrix, and St=Gt1/2S_t=G_t^{1/2}St​=Gt1/2​ is its positive semidefinite square root. With parameters η>0\eta>0η>0 and δ≥0\delta\ge0δ≥0, ADAGRAD with full matrices (Figure 2 of the paper) sets

Ht=δI+St,ψt(x)=12⟨x,Htx⟩,Bψt(x,y)=12⟨x−y,Ht(x−y)⟩,H_t=\delta I+S_t,\qquad \psi_t(x)=\tfrac12\langle x,H_tx\rangle,\qquad B_{\psi_t}(x,y)=\tfrac12\langle x-y,H_t(x-y)\rangle,Ht​=δI+St​,ψt​(x)=21​⟨x,Ht​x⟩,Bψt​​(x,y)=21​⟨x−y,Ht​(x−y)⟩,

starts at x1=0x_1=0x1​=0, and computes xt+1x_{t+1}xt+1​ by one of two updates:

  • the primal-dual subgradient update (3): xt+1∈argmin⁡x∈X{η⟨1t∑τ≤tgτ,x⟩+ηφ(x)+1tψt(x)}x_{t+1}\in\operatorname{argmin}_{x\in\mathcal X}\{\eta\langle\tfrac1t\sum_{\tau\le t}g_\tau,x\rangle+\eta\varphi(x)+\tfrac1t\psi_t(x)\}xt+1​∈argminx∈X​{η⟨t1​∑τ≤t​gτ​,x⟩+ηφ(x)+t1​ψt​(x)};
  • the composite mirror descent update (4): xt+1∈argmin⁡x∈X{η⟨gt,x⟩+ηφ(x)+Bψt(x,xt)}x_{t+1}\in\operatorname{argmin}_{x\in\mathcal X}\{\eta\langle g_t,x\rangle+\eta\varphi(x)+B_{\psi_t}(x,x_t)\}xt+1​∈argminx∈X​{η⟨gt​,x⟩+ηφ(x)+Bψt​​(x,xt​)}.

The dual (semi)norm of ψt\psi_tψt​ is ∥v∥ψt∗2=⟨v,Ht†v⟩\|v\|_{\psi_t^*}^2=\langle v,H_t^\dagger v\rangle∥v∥ψt∗​2​=⟨v,Ht†​v⟩, where A†A^\daggerA† is the pseudo-inverse.

Formalization targets

Goal: Theorem 7

Assume X\mathcal XX closed and convex, ftf_tft​ and φ\varphiφ convex, and φ\varphiφ minimized over X\mathcal XX at x1=0x_1=0x1​=0. For the primal-dual update with δ≥max⁡t≤T∥gt∥2\delta\ge\max_{t\le T}\|g_t\|_2δ≥maxt≤T​∥gt​∥2​, every x∗∈Xx^*\in\mathcal Xx∗∈X satisfies

Rϕ(T)≤δη∥x∗∥22+1η∥x∗∥22tr⁡(GT1/2)+ηtr⁡(GT1/2),R_\phi(T)\le\frac\delta\eta\|x^*\|_2^2+\frac1\eta\|x^*\|_2^2\operatorname{tr}(G_T^{1/2})+\eta\operatorname{tr}(G_T^{1/2}),Rϕ​(T)≤ηδ​∥x∗∥22​+η1​∥x∗∥22​tr(GT1/2​)+ηtr(GT1/2​),

and for the composite mirror descent update with any δ≥0\delta\ge0δ≥0,

Rϕ(T)≤δη∥x∗∥22+12ηmax⁡t≤T∥x∗−xt∥22tr⁡(GT1/2)+ηtr⁡(GT1/2).R_\phi(T)\le\frac\delta\eta\|x^*\|_2^2+\frac1{2\eta}\max_{t\le T}\|x^*-x_t\|_2^2\operatorname{tr}(G_T^{1/2})+\eta\operatorname{tr}(G_T^{1/2}).Rϕ​(T)≤ηδ​∥x∗∥22​+2η1​t≤Tmax​∥x∗−xt​∥22​tr(GT1/2​)+ηtr(GT1/2​).

Milestones

In the order the proof uses them:

  1. Lemma 16 and Propositions 3 and 2: regret bounds for mirror descent and dual averaging with time-varying quadratic proximal functions ψt(x)=12⟨x,Htx⟩\psi_t(x)=\tfrac12\langle x,H_tx\rangleψt​(x)=21​⟨x,Ht​x⟩ ((11) and (10)).
  2. Lemma 13: A⪰B⪰0A\succeq B\succeq0A⪰B⪰0 implies A1/2⪰B1/2A^{1/2}\succeq B^{1/2}A1/2⪰B1/2.
  3. The first display of the proof of Theorem 7 and inequality (16): the growth of the Bregman terms is controlled by tr⁡(Gt1/2)\operatorname{tr}(G_t^{1/2})tr(Gt1/2​).
  4. Lemma 14 (∇Xtr⁡(Xp)=pXp−1\nabla_X\operatorname{tr}(X^p)=pX^{p-1}∇X​tr(Xp)=pXp−1), Lemma 8 (a first-order concavity inequality for tr⁡(B1/2)\operatorname{tr}(B^{1/2})tr(B1/2) at singular BBB), Lemma 9, and Lemma 10, the doubling lemma
∑t=1T⟨gt,St†gt⟩≤2∑t=1T⟨gt,ST†gt⟩=2tr⁡(GT1/2).\sum_{t=1}^T\langle g_t,S_t^\dagger g_t\rangle\le2\sum_{t=1}^T\langle g_t,S_T^\dagger g_t\rangle=2\operatorname{tr}(G_T^{1/2}).t=1∑T​⟨gt​,St†​gt​⟩≤2t=1∑T​⟨gt​,ST†​gt​⟩=2tr(GT1/2​).
  1. (17): the dual-norm sums of both updates are at most 2tr⁡(GT1/2)2\operatorname{tr}(G_T^{1/2})2tr(GT1/2​).

Significance

The quantity tr⁡(GT1/2)\operatorname{tr}(G_T^{1/2})tr(GT1/2​) can be much smaller than the Tmax⁡t∥gt∥2\sqrt{T}\max_t\|g_t\|_2T​maxt​∥gt​∥2​ factor in the regret of non-adaptive online gradient descent with a tuned step size: it is small when the subgradients concentrate in a few directions, possibly not aligned with the coordinate axes. The paper's Corollary 11 rewrites it as d\sqrt dd​ times the square root of inf⁡{∑tgt⊤S−1gt:S⪰0,tr⁡(S)≤d}\inf\{\sum_t g_t^\top S^{-1}g_t: S\succeq0,\operatorname{tr}(S)\le d\}inf{∑t​gt⊤​S−1gt​:S⪰0,tr(S)≤d}, the gradient term of the best fixed full-matrix proximal function chosen in hindsight. The supporting results have their own uses. Propositions 2 and 3 are the generic regret bounds for adaptive proximal methods. Lemmas 8–10 are trace inequalities for the matrix square root that appear throughout the analysis of second-order online methods.

The results are proved in the paper. To our knowledge none of them has a machine-checked proof. Formalizing them requires the matrix functional calculus (square roots, pseudo-inverses, real powers) in a form usable for inequalities, which Mathlib has only partly: for instance, monotonicity of the square root is in Mathlib for C*-algebras, which covers complex but not real matrices. A complete development gives a verified regret bound for an adaptive full-matrix method together with these matrix inequalities.

Difficulty

The online-learning half (Lemma 16, Propositions 2 and 3) is a convex-analysis argument once the proximal functions are quadratic. The obstacle is the matrix half. The natural approach to Lemma 10 is induction with a scalar inequality b−a≤b−a/(2b)\sqrt{b-a}\le\sqrt b-a/(2\sqrt b)b−a​≤b​−a/(2b​) applied eigenvalue by eigenvalue. This fails because GT−1G_{T-1}GT−1​ and GTG_TGT​ do not commute, so their eigenvectors differ. One needs the concavity of A↦tr⁡(A1/2)A\mapsto\operatorname{tr}(A^{1/2})A↦tr(A1/2) on positive semidefinite matrices and its gradient, at points where AAA may be singular. That is where Lemma 8 and the pseudo-inverse enter: the inverse root does not exist, and a limit δ↓0\delta\downarrow0δ↓0 is needed. The same issue arises in Lemma 9 and in the dual norm when δ=0\delta=0δ=0.

Formalization scope

Vectors are EuclideanSpace ℝ (Fin d), so ‖·‖ is the Euclidean norm. Matrices are Matrix (Fin d) (Fin d) ℝ, acting on vectors through Matrix.toEuclideanLin. The square root is Mathlib's CFC.sqrt under the Loewner order (MatrixOrder). The pseudo-inverse is the functional calculus of λ↦λ−1\lambda\mapsto\lambda^{-1}λ↦λ−1 with 0−1=00^{-1}=00−1=0, and B−1/2B^{-1/2}B−1/2 is written as (B†)1/2(B^\dagger)^{1/2}(B†)1/2. Mathlib's matrix inverse is used only where the matrix is invertible or the vector is zero. Rounds are 1-based and G0=0G_0=0G0​=0. The losses and φ\varphiφ are real-valued convex functions, and the constraint is carried by X\mathcal XX. Each update's argmin is a predicate (the next iterate lies in X\mathcal XX and minimizes the objective there), which does not assert existence or uniqueness. A run consists of the TTT rounds of Figure 2, with the subgradient relation imposed in every round. The subgradient relation is the platform's published ShorNonsmooth.AlmostDiff.IsSubgradient.

Restrictions and conventions, each stated in the item that needs it:

  • φ\varphiφ is minimized over X\mathcal XX at x1=0x_1=0x1​=0. This is the paper's x1=arg⁡min⁡Xφx_1=\arg\min_{\mathcal X}\varphix1​=argminX​φ. Without it both parts of Theorem 7 are false.
  • Lemma 8 assumes ν≥0\nu\ge0ν≥0. The page says "for any ν\nuν", but the inequality fails for ν<0\nu<0ν<0 (for B=0B=0B=0, ν=−1\nu=-1ν=−1 it reads 2∥g∥2≤02\|g\|_2\le02∥g∥2​≤0). The paper only applies the lemma with ν=1\nu=1ν=1.
  • Lemma 16 and Propositions 2, 3 are specialised to quadratic proximal functions. In Lemma 16 and Proposition 3, HtH_tHt​ is positive semidefinite and gtg_tgt​ lies in its range, so the dual norm is finite. In Proposition 2, HtH_tHt​ is positive definite, the HtH_tHt​ increase, and x1=0x_1=0x1​=0. Lemma 16 asks for x∗∈Xx^*\in\mathcal Xx∗∈X. Proposition 3 is stated for T≥1T\ge1T≥1 rounds.
  • H0=δIH_0=\delta IH0​=δI in the dual norm ∥⋅∥ψ0∗\|\cdot\|_{\psi_0^*}∥⋅∥ψ0∗​​ of (17) and Theorem 7. Figure 2's H0=0H_0=0H0​=0 is never used by the updates.
  • Lemma 14's gradient is stated as a directional derivative along every symmetric direction.

A formalization in which the next iterate need not lie in X\mathcal XX, the comparator ranges outside X\mathcal XX, StS_tSt​ is GtG_tGt​ itself or its entrywise square root, or St†S_t^\daggerSt†​ is Mathlib's matrix inverse (which vanishes on singular matrices), states a different theorem and is excluded.

Proofs need the convex-analysis optimality conditions for the two updates, the Loewner-order calculus of CFC.sqrt on real matrices (Lemma 13), concavity and differentiability of tr⁡(A1/2)\operatorname{tr}(A^{1/2})tr(A1/2), and limits δ↓0\delta\downarrow0δ↓0 in the functional calculus. The matrix lemmas (13, 14, 8, 9, 10) are reusable outside this paper. Alternative proofs, for instance of Lemma 10 through operator concavity, are welcome.

Selected references

  • J. Duchi, E. Hazan, Y. Singer, Adaptive Subgradient Methods for Online Learning and Stochastic Optimization, Journal of Machine Learning Research 12 (2011) 2121–2159. https://jmlr.org/papers/v12/duchi11a.html
  • H. B. McMahan, M. Streeter, Adaptive Bound Optimization for Online Convex Optimization, COLT 2010. https://arxiv.org/abs/1002.4908
  • L. Xiao, Dual Averaging Methods for Regularized Stochastic Learning and Online Optimization, Journal of Machine Learning Research 11 (2010) 2543–2596. https://jmlr.org/papers/v11/xiao10a.html
  • Y. Nesterov, Primal-dual subgradient methods for convex problems, Mathematical Programming 120 (2009) 221–259. https://doi.org/10.1007/s10107-007-0149-x
  • T. Ando, Concavity of certain maps on positive definite matrices and applications to Hadamard products, Linear Algebra and its Applications 26 (1979) 203–241. https://doi.org/10.1016/0024-3795(79)90179-4
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OptimizationReinforcement Learning·Captain: mikedeng1

On the Theory of Policy Gradient Methods: Optimality, Approximation, and Distribution Shift 5: Natural Policy Gradient with Softmax Policies Is Within log|A|/(ηT) + 1/((1−γ)²T) of OptimalResearch Paper

Why natural policy gradient

Policy gradient methods optimize a parameterized policy directly by gradient ascent on its expected discounted return. They underlie most practical reinforcement-learning systems, yet for a long time their convergence was understood only locally: the objective θ↦Vπθ(ρ)\theta\mapsto V^{\pi_\theta}(\rho)θ↦Vπθ​(ρ) is not concave, and plain gradient ascent can stall on plateaus. The natural policy gradient (NPG) of Kakade (2001) preconditions the gradient by the pseudoinverse of the Fisher information matrix of the policy, so that steps are measured in the geometry of the induced action distributions rather than in parameter space. NPG is the conceptual ancestor of trust-region methods such as TRPO and PPO.

Agarwal, Kakade, Lee and Mahajan (arXiv:1908.00261v5, JMLR 2021) prove that exact NPG with softmax policies converges to the optimal value at rate O(1/T)O(1/T)O(1/T), with constants that depend on neither the number of states, nor the number of actions beyond log⁡∣A∣\log|\mathcal A|log∣A∣, nor any distribution mismatch coefficient. This mission formalizes that result, Theorem 5.3 of the paper, and the lemmas its proof rests on.

Timeline. Kakade (2001) introduced NPG and observed that, for softmax policies, it moves toward the greedy policy. Kakade and Langford (2002) proved the performance difference lemma. Even-Dar, Kakade and Mansour (2009) analyzed experts algorithms in changing MDPs, whose regret argument this proof adapts; Geist, Scherrer and Pietquin (2019) studied regularized policy iteration with related rates. Agarwal et al. (preprint 2019, journal 2021) gave the dimension-free log⁡∣A∣/(ηT)+1/((1−γ)2T)\log|\mathcal A|/(\eta T)+1/((1-\gamma)^2T)log∣A∣/(ηT)+1/((1−γ)2T) bound formalized here.

Setting

A finite discounted MDP consists of finite sets S\mathcal SS (states) and A\mathcal AA (actions, nonempty), a transition kernel P(s′∣s,a)P(s'\mid s,a)P(s′∣s,a), rewards r(s,a)∈[0,1]r(s,a)\in[0,1]r(s,a)∈[0,1] and a discount factor γ∈[0,1)\gamma\in[0,1)γ∈[0,1). A policy π\piπ assigns to each state a distribution π(⋅∣s)\pi(\cdot\mid s)π(⋅∣s) over actions. Its value Vπ(s)V^\pi(s)Vπ(s) is the expected discounted sum of rewards started from sss; for a start distribution ρ\rhoρ, Vπ(ρ)=∑sρ(s)Vπ(s)V^\pi(\rho)=\sum_s\rho(s)V^\pi(s)Vπ(ρ)=∑s​ρ(s)Vπ(s). The state–action value is Qπ(s,a)=r(s,a)+γ∑s′P(s′∣s,a)Vπ(s′)Q^\pi(s,a)=r(s,a)+\gamma\sum_{s'}P(s'\mid s,a)V^\pi(s')Qπ(s,a)=r(s,a)+γ∑s′​P(s′∣s,a)Vπ(s′) and the advantage is Aπ(s,a)=Qπ(s,a)−Vπ(s)A^\pi(s,a)=Q^\pi(s,a)-V^\pi(s)Aπ(s,a)=Qπ(s,a)−Vπ(s). The discounted state visitation distribution is

dρπ(s)=(1−γ)∑t≥0γt Pr⁡π(st=s∣s0∼ρ).d^\pi_\rho(s)=(1-\gamma)\sum_{t\ge0}\gamma^t\,\Pr{}^\pi(s_t=s\mid s_0\sim\rho).dρπ​(s)=(1−γ)t≥0∑​γtPrπ(st​=s∣s0​∼ρ).

An optimal policy π⋆\pi^\starπ⋆ has Vπ⋆(s)≥Vπ(s)V^{\pi^\star}(s)\ge V^\pi(s)Vπ⋆(s)≥Vπ(s) for every policy π\piπ and every state sss; write V⋆=Vπ⋆V^\star=V^{\pi^\star}V⋆=Vπ⋆.

The softmax policy with parameter θ∈R∣S∣∣A∣\theta\in\mathbb R^{|\mathcal S||\mathcal A|}θ∈R∣S∣∣A∣ is πθ(a∣s)=exp⁡(θs,a)/∑a′exp⁡(θs,a′)\pi_\theta(a\mid s)=\exp(\theta_{s,a})/\sum_{a'}\exp(\theta_{s,a'})πθ​(a∣s)=exp(θs,a​)/∑a′​exp(θs,a′​). The Fisher information matrix with respect to ρ\rhoρ is

Fρ(θ)=Es∼dρπθEa∼πθ(⋅∣s)[∇θlog⁡πθ(a∣s) ∇θlog⁡πθ(a∣s)⊤],F_\rho(\theta)=\mathbb E_{s\sim d^{\pi_\theta}_\rho}\mathbb E_{a\sim\pi_\theta(\cdot\mid s)}\big[\nabla_\theta\log\pi_\theta(a\mid s)\,\nabla_\theta\log\pi_\theta(a\mid s)^\top\big],Fρ​(θ)=Es∼dρπθ​​​Ea∼πθ​(⋅∣s)​[∇θ​logπθ​(a∣s)∇θ​logπθ​(a∣s)⊤],

and the NPG update with step size η\etaη is

θ(t+1)=θ(t)+η Fρ(θ(t))† ∇θV(t)(ρ),(15)\theta^{(t+1)}=\theta^{(t)}+\eta\,F_\rho(\theta^{(t)})^\dagger\,\nabla_\theta V^{(t)}(\rho),\tag{15}θ(t+1)=θ(t)+ηFρ​(θ(t))†∇θ​V(t)(ρ),(15)

where M†M^\daggerM† is the Moore–Penrose pseudoinverse and π(t),V(t),A(t)\pi^{(t)},V^{(t)},A^{(t)}π(t),V(t),A(t) are the quantities of θ(t)\theta^{(t)}θ(t). Following §5.3, every state is assumed reachable from ρ\rhoρ: for each sss some policy π\piπ has dρπ(s)>0d^\pi_\rho(s)>0dρπ​(s)>0. The normalizer Zt(s)=∑aπ(t)(a∣s)exp⁡(ηA(t)(s,a)/(1−γ))Z_t(s)=\sum_a\pi^{(t)}(a\mid s)\exp(\eta A^{(t)}(s,a)/(1-\gamma))Zt​(s)=∑a​π(t)(a∣s)exp(ηA(t)(s,a)/(1−γ)) appears in the lemmas.

Formalization targets

Goal: Theorem 5.3

For NPG (15) from θ(0)=0\theta^{(0)}=0θ(0)=0 with any fixed η>0\eta>0η>0, for all T>0T>0T>0,

V(T)(ρ) ≥ V⋆(ρ)−log⁡∣A∣ηT−1(1−γ)2T.V^{(T)}(\rho)\ \ge\ V^\star(\rho)-\frac{\log|\mathcal A|}{\eta T}-\frac{1}{(1-\gamma)^2T}.V(T)(ρ) ≥ V⋆(ρ)−ηTlog∣A∣​−(1−γ)2T1​.

Companion: iteration complexity

If moreover η≥(1−γ)2log⁡∣A∣\eta\ge(1-\gamma)^2\log|\mathcal A|η≥(1−γ)2log∣A∣ and ϵ>0\epsilon>0ϵ>0, then V⋆(ρ)−V(T)(ρ)≤ϵV^\star(\rho)-V^{(T)}(\rho)\le\epsilonV⋆(ρ)−V(T)(ρ)≤ϵ for every T≥2/((1−γ)2ϵ)T\ge 2/((1-\gamma)^2\epsilon)T≥2/((1−γ)2ϵ).

Milestones

  1. Lemma 3.2 (performance difference): Vπ(s0)−Vπ′(s0)=11−γEs∼ds0πEa∼π(⋅∣s)[Aπ′(s,a)]V^\pi(s_0)-V^{\pi'}(s_0)=\frac{1}{1-\gamma}\mathbb E_{s\sim d^\pi_{s_0}}\mathbb E_{a\sim\pi(\cdot\mid s)}[A^{\pi'}(s,a)]Vπ(s0​)−Vπ′(s0​)=1−γ1​Es∼ds0​π​​Ea∼π(⋅∣s)​[Aπ′(s,a)].
  2. Lemma 5.1 (soft policy iteration, policy form): π(t+1)(a∣s)=π(t)(a∣s)exp⁡(ηA(t)(s,a)/(1−γ))/Zt(s)\pi^{(t+1)}(a\mid s)=\pi^{(t)}(a\mid s)\exp(\eta A^{(t)}(s,a)/(1-\gamma))/Z_t(s)π(t+1)(a∣s)=π(t)(a∣s)exp(ηA(t)(s,a)/(1−γ))/Zt​(s).
  3. Lemma 5.2 (improvement): V(t+1)(μ)−V(t)(μ)≥1−γηEs∼μlog⁡Zt(s)≥0V^{(t+1)}(\mu)-V^{(t)}(\mu)\ge\frac{1-\gamma}{\eta}\mathbb E_{s\sim\mu}\log Z_t(s)\ge0V(t+1)(μ)−V(t)(μ)≥η1−γ​Es∼μ​logZt​(s)≥0 for every μ∈Δ(S)\mu\in\Delta(\mathcal S)μ∈Δ(S).
  4. Monotonicity (proof of Theorem 5.3): V(t+1)(s)≥V(t)(s)V^{(t+1)}(s)\ge V^{(t)}(s)V(t+1)(s)≥V(t)(s) for all sss.

Significance

The result. Theorem 5.3 is a global, non-asymptotic guarantee for a non-concave policy optimization problem, with no dependence on ∣S∣|\mathcal S|∣S∣ and no distribution mismatch coefficient: the same analysis does not go through for plain softmax policy gradient, whose rates in the same paper carry such coefficients or are only asymptotic. It explains why Fisher preconditioning helps, and it is the template for the paper's later analysis of NPG with function approximation (§6) and for subsequent work on linear convergence of NPG and on entropy-regularized variants.

Formalizing it. The theorem is proved on paper; no machine-checked proof is known. A formal development needs: a policy-gradient identity for softmax policies, the computation of the Moore–Penrose pseudoinverse of a singular, block-structured Fisher matrix acting on the policy gradient, the performance difference lemma in the unnormalized convention, and a telescoping regret argument with Kullback–Leibler divergences between action distributions. Each of these is reusable: the performance difference lemma and the visitation-distribution calculus serve every result of the paper and much of the reinforcement-learning literature.

Difficulty

The obvious approach, treating NPG as gradient ascent on a smooth objective, fails: Vπθ(ρ)V^{\pi_\theta}(\rho)Vπθ​(ρ) is non-concave and its smoothness constant grows with 1/(1−γ)31/(1-\gamma)^31/(1−γ)3, which would give a rate depending on the state space and on distribution mismatch. The dimension-free rate comes only after the NPG step is rewritten exactly as a multiplicative-weights update on each state's action distribution, and this rewriting is itself delicate. The Fisher matrix is always singular (each state's block annihilates the all-ones vector), the pseudoinverse returns the minimum-norm solution, which is not A(t)/(1−γ)A^{(t)}/(1-\gamma)A(t)/(1−γ) but differs from it by a state-dependent offset; and the identity holds at a state only if that state carries positive visitation weight, which is why reachability from ρ\rhoρ is assumed. A newcomer's first idea, replacing F†F^\daggerF† by "any solution of Fu=∇VFu=\nabla VFu=∇V" or by an ordinary inverse, does not define the algorithm.

Formalization scope

Lean conventions (Mathlib, namespace PolicyGradTheory.NPG):

  • Policies are functions S → A → ℝ (π s a =π(a∣s)=\pi(a\mid s)=π(a∣s)) satisfying the published IsPolicy; transitions P s a s' =P(s′∣s,a)=P(s'\mid s,a)=P(s′∣s,a); values, QQQ-functions and state occupation probabilities are the published FoundationsML definitions PolicyValue, QFunction, OccupationDist, and optimality is IsOptimalPolicy. V⋆(ρ)V^\star(\rho)V⋆(ρ) is the value of a policy assumed optimal, never a supremum over arbitrary functions.
  • Parameters live in EuclideanSpace ℝ (S × A), so norms are Euclidean and gradient is ∇θ\nabla_\theta∇θ​.
  • The pseudoinverse is the published predicate VectorSpaceOpt.IsPseudoinverse (minimum-norm least-squares inverse, which on Euclidean space is the Moore–Penrose pseudoinverse); Mathlib has none.
  • Standing assumptions carried by every statement: rewards in [0,1][0,1][0,1], γ∈[0,1)\gamma\in[0,1)γ∈[0,1), A\mathcal AA nonempty, ρ∈Δ(S)\rho\in\Delta(\mathcal S)ρ∈Δ(S), every state reachable from ρ\rhoρ (§5.3, footnote 6). The goal also assumes θ(0)=0\theta^{(0)}=0θ(0)=0 and η>0\eta>0η>0, as the page does. Lemmas 5.1 and 5.2 hold from any start.
  • log⁡\loglog is the natural logarithm; TTT counts updates and T>0T>0T>0 is a natural number.

The goal does not assume the soft-policy-iteration form, the improvement bound or monotonicity; those are milestones to be proved, and a statement taking any of them as a hypothesis would not be Theorem 5.3. The parameter form "θ(t+1)=θ(t)+η1−γA(t)\theta^{(t+1)}=\theta^{(t)}+\frac{\eta}{1-\gamma}A^{(t)}θ(t+1)=θ(t)+1−γη​A(t)" of Lemma 5.1 is not stated, since the pseudoinverse step differs from it by a state-dependent offset (as the lemma's own proof on p. 70 shows); only the policy form, which is invariant to that offset, is a target.

Contributions welcome: the softmax policy-gradient identity and score-function formula, existence and uniqueness of the minimum-norm least-squares inverse on Euclidean space, Kullback–Leibler identities for finite distributions, and the summability facts behind the visitation distribution.

Selected references

  • A. Agarwal, S. M. Kakade, J. D. Lee, G. Mahajan, On the Theory of Policy Gradient Methods: Optimality, Approximation, and Distribution Shift, JMLR 22(98), 2021; preprint arXiv:1908.00261v5, 2020. https://arxiv.org/abs/1908.00261v5
  • S. M. Kakade, A Natural Policy Gradient, NeurIPS 2001. https://proceedings.neurips.cc/paper/2001/hash/4b86abe48d358ecf194c56c69108433e-Abstract.html
  • S. Kakade, J. Langford, Approximately Optimal Approximate Reinforcement Learning, ICML 2002. https://dl.acm.org/doi/10.5555/645531.656005
  • E. Even-Dar, S. M. Kakade, Y. Mansour, Online Markov Decision Processes, Mathematics of Operations Research 34(3), 2009. https://doi.org/10.1287/moor.1090.0396
  • M. Geist, B. Scherrer, O. Pietquin, A Theory of Regularized Markov Decision Processes, ICML 2019. https://arxiv.org/abs/1901.11275
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Convex OptimizationOptimization·Captain: mikedeng1

Adaptive Subgradient Methods for Online Learning and Stochastic Optimization 1: Diagonal AdaGrad's Regret Is Bounded by the Per-Coordinate Gradient Norms Σᵢ‖g_{1:T,i}‖₂Research Paper

Motivation

Stochastic and online subgradient methods are the workhorse of large-scale learning: each step touches one example, costs time linear in the dimension, and needs no line search. Their weak point is the step size. A single global step size treats every coordinate alike, while in high-dimensional, sparse problems (text, click-through data, bag-of-words features) most coordinates are zero in most examples, and the few informative rare features receive vanishingly small updates.

AdaGrad (Duchi, Hazan and Singer, JMLR 2011; McMahan and Streeter, COLT 2010, independently) chooses a separate step size per coordinate from the gradients observed so far. Its diagonal version is the ancestor of RMSProp and Adam and is one of the most widely used optimizers in machine learning. This mission targets the paper's regret guarantee for diagonal AdaGrad, which explains when per-coordinate adaptation pays off.

Timeline. Zinkevich (2003) proved O(T)O(\sqrt T)O(T​) regret for online projected gradient descent. Nesterov (2009) and Xiao (2010) analysed dual averaging with a regularizer; Duchi, Shalev-Shwartz, Singer and Tewari (2010) analysed composite mirror descent. Auer and Gentile (2000) proved the scalar inequality that makes adaptive step sizes work. Duchi, Hazan and Singer (2011) combined these into adaptive proximal functions with data-dependent regret bounds.

Setting

Work in Rd\mathbb R^dRd with the Euclidean inner product and ∥x∥∞=max⁡i∣xi∣\|x\|_\infty=\max_i|x_i|∥x∥∞​=maxi​∣xi​∣. A closed convex set X⊆Rd\mathcal X\subseteq\mathbb R^dX⊆Rd with 0∈X0\in\mathcal X0∈X is fixed, together with a convex regularizer φ:Rd→R\varphi:\mathbb R^d\to\mathbb Rφ:Rd→R (for example λ∥x∥1\lambda\|x\|_1λ∥x∥1​). In rounds t=1,2,…t=1,2,\dotst=1,2,… a learner plays xt∈Xx_t\in\mathcal Xxt​∈X, a convex loss ftf_tft​ is revealed, and the learner observes a subgradient gtg_tgt​ of ftf_tft​ at xtx_txt​: ft(y)≥ft(xt)+⟨gt,y−xt⟩f_t(y)\ge f_t(x_t)+\langle g_t,y-x_t\rangleft​(y)≥ft​(xt​)+⟨gt​,y−xt​⟩ for all yyy. The regret against a comparator x∗∈Xx^*\in\mathcal Xx∗∈X is

Rφ(T)=∑t=1T[ft(xt)+φ(xt)−ft(x∗)−φ(x∗)].R_\varphi(T)=\sum_{t=1}^T\big[f_t(x_t)+\varphi(x_t)-f_t(x^*)-\varphi(x^*)\big].Rφ​(T)=t=1∑T​[ft​(xt​)+φ(xt​)−ft​(x∗)−φ(x∗)].

For each coordinate iii, let g1:t,i=(g1,i,…,gt,i)g_{1:t,i}=(g_{1,i},\dots,g_{t,i})g1:t,i​=(g1,i​,…,gt,i​) and st,i=∥g1:t,i∥2s_{t,i}=\|g_{1:t,i}\|_2st,i​=∥g1:t,i​∥2​. Diagonal AdaGrad (Figure 1 of the paper) has parameters η>0\eta>0η>0, δ≥0\delta\ge0δ≥0, starts at x1=0x_1=0x1​=0, and in round ttt uses the proximal function ψt(x)=12⟨x,(δI+diag(st))x⟩\psi_t(x)=\frac12\langle x,(\delta I+\mathrm{diag}(s_t))x\rangleψt​(x)=21​⟨x,(δI+diag(st​))x⟩. It updates by one of two rules:

  • the primal-dual subgradient update (3): xt+1x_{t+1}xt+1​ minimizes η⟨1t∑τ≤tgτ,x⟩+ηφ(x)+1tψt(x)\eta\langle\frac1t\sum_{\tau\le t}g_\tau,x\rangle+\eta\varphi(x)+\frac1t\psi_t(x)η⟨t1​∑τ≤t​gτ​,x⟩+ηφ(x)+t1​ψt​(x) over X\mathcal XX;
  • the composite mirror descent update (4): xt+1x_{t+1}xt+1​ minimizes η⟨gt,x⟩+ηφ(x)+Bψt(x,xt)\eta\langle g_t,x\rangle+\eta\varphi(x)+B_{\psi_t}(x,x_t)η⟨gt​,x⟩+ηφ(x)+Bψt​​(x,xt​) over X\mathcal XX, where Bψt(x,y)=12∑i(δ+st,i)(xi−yi)2B_{\psi_t}(x,y)=\frac12\sum_i(\delta+s_{t,i})(x_i-y_i)^2Bψt​​(x,y)=21​∑i​(δ+st,i​)(xi​−yi​)2 is the Bregman divergence of ψt\psi_tψt​.

Formalization targets

Goal: Theorem 5

For the primal-dual update with δ≥max⁡t≤T∥gt∥∞\delta\ge\max_{t\le T}\|g_t\|_\inftyδ≥maxt≤T​∥gt​∥∞​, and every x∗∈Xx^*\in\mathcal Xx∗∈X,

Rφ(T)≤δη∥x∗∥22+1η∥x∗∥∞2∑i=1d∥g1:T,i∥2+η∑i=1d∥g1:T,i∥2.R_\varphi(T)\le\frac\delta\eta\|x^*\|_2^2+\frac1\eta\|x^*\|_\infty^2\sum_{i=1}^d\|g_{1:T,i}\|_2+\eta\sum_{i=1}^d\|g_{1:T,i}\|_2 .Rφ​(T)≤ηδ​∥x∗∥22​+η1​∥x∗∥∞2​i=1∑d​∥g1:T,i​∥2​+ηi=1∑d​∥g1:T,i​∥2​.

For the composite mirror descent update with δ=0\delta=0δ=0, and every x∗∈Xx^*\in\mathcal Xx∗∈X,

Rφ(T)≤12ηmax⁡t≤T∥x∗−xt∥∞2∑i=1d∥g1:T,i∥2+η∑i=1d∥g1:T,i∥2.R_\varphi(T)\le\frac1{2\eta}\max_{t\le T}\|x^*-x_t\|_\infty^2\sum_{i=1}^d\|g_{1:T,i}\|_2+\eta\sum_{i=1}^d\|g_{1:T,i}\|_2 .Rφ​(T)≤2η1​t≤Tmax​∥x∗−xt​∥∞2​i=1∑d​∥g1:T,i​∥2​+ηi=1∑d​∥g1:T,i​∥2​.

Milestones

  1. Lemma 16: the one-step inequality of composite mirror descent.
  2. Proposition 3: the regret of composite mirror descent with time-varying proximal functions.
  3. Proposition 2: the regret of primal-dual subgradient (dual averaging) with time-varying proximal functions.
  4. Inequality (24): ∑t≤Tat2/∥a1:t∥2≤2∥a1:T∥2\sum_{t\le T}a_t^2/\|a_{1:t}\|_2\le2\|a_{1:T}\|_2∑t≤T​at2​/∥a1:t​∥2​≤2∥a1:T​∥2​ for any real sequence.
  5. Lemma 4: ∑t≤T⟨gt,diag(st)−1gt⟩≤2∑i∥g1:T,i∥2\sum_{t\le T}\langle g_t,\mathrm{diag}(s_t)^{-1}g_t\rangle\le2\sum_i\|g_{1:T,i}\|_2∑t≤T​⟨gt​,diag(st​)−1gt​⟩≤2∑i​∥g1:T,i​∥2​.
  6. Inequality (13): the mirror-descent gradient term is at most 2∑i∥g1:T,i∥22\sum_i\|g_{1:T,i}\|_22∑i​∥g1:T,i​∥2​.
  7. The primal-dual analogue of (13) (§3, after (13)).
  8. Inequality (14): the drift of the Bregman divergences is controlled by max⁡t∥x∗−xt∥∞2∑i∥g1:T,i∥2\max_t\|x^*-x_t\|_\infty^2\sum_i\|g_{1:T,i}\|_2maxt​∥x∗−xt​∥∞2​∑i​∥g1:T,i​∥2​.

Lemma 16 and Propositions 2 and 3 are stated in the paper for general proximal functions; here they are specialised to diagonal quadratic ones, ψt(y)=12∑iht,iyi2\psi_t(y)=\frac12\sum_ih_{t,i}y_i^2ψt​(y)=21​∑i​ht,i​yi2​.

Significance

The result. The quantity ∑i∥g1:T,i∥2\sum_i\|g_{1:T,i}\|_2∑i​∥g1:T,i​∥2​ is never larger than d (∑t∥gt∥22)1/2\sqrt d\,\big(\sum_t\|g_t\|_2^2\big)^{1/2}d​(∑t​∥gt​∥22​)1/2 and can be far smaller when gradients are sparse or coordinate scales differ. Theorem 5 shows that diagonal AdaGrad does as well as the best fixed diagonal preconditioner chosen in hindsight, up to a constant, without knowing the gradients in advance (Corollary 1 and Corollary 6 of the paper). Through online-to-batch conversion the same bound gives convergence rates for stochastic convex optimization with sparse data. The theorem underpins the theoretical case for per-coordinate adaptive step sizes.

Formalizing it. The result is proved on paper; to our knowledge no machine-checked proof of AdaGrad's regret bound exists, in Lean or elsewhere. The formalization produces checked versions of the two regret templates (composite mirror descent and dual averaging with changing proximal functions), which are reused by most later analyses of adaptive methods, and of the Auer–Gentile inequality. The formal statements also make precise two hypotheses the printed theorem leaves implicit (see Formalization scope).

Difficulty

The obvious argument fixes one proximal function and applies the classical mirror-descent or dual-averaging bound. That fails because AdaGrad's ψt\psi_tψt​ changes every round and depends on the very subgradients whose norms are being bounded: the classical bounds assume a fixed ψ\psiψ, and with changing ψt\psi_tψt​ new drift terms appear that must be shown to be small. The gradient term ∑t∥gt∥ψt∗2\sum_t\|g_t\|^2_{\psi_t^*}∑t​∥gt​∥ψt∗​2​ has denominators that grow with the data, so it is not bounded term by term; it has to be bounded as a whole sum. For dual averaging there is an additional index shift: round ttt's subgradient is measured in the dual norm of ψt−1\psi_{t-1}ψt−1​, which has not yet seen gtg_tgt​.

Formalization scope

Vectors live in EuclideanSpace ℝ (Fin d), so ‖·‖ is the Euclidean norm; ∥⋅∥∞\|\cdot\|_\infty∥⋅∥∞​ is a separate definition supNorm. Rounds are 1-based, with sums over Finset.Icc 1 T; s0=0s_0=0s0​=0. Losses and the regularizer are real-valued convex functions on all of Rd\mathbb R^dRd, so extended-valued regularizers (indicator functions) are excluded, and the constraint is carried by X\mathcal XX. The subgradient relation is the published platform definition ShorNonsmooth.AlmostDiff.IsSubgradient. Each update is a predicate (xt+1∈Xx_{t+1}\in\mathcal Xxt+1​∈X and minimizes the update objective over X\mathcal XX), since the minimizer need not be unique; the theorems hold for every run. The factor 12\frac1221​ in ψt\psi_tψt​ follows Figure 1. Dual norms use Lean's division, where a/0=0a/0=0a/0=0 is the paper's 0/0=00/0=00/0=0. The losses are a fixed sequence, which also covers adaptive adversaries because the run is determined by the losses.

Two restrictions relative to the printed Theorem 5, both forced by the paper's own proof:

  1. The mirror-descent part is stated for δ=0\delta=0δ=0. The proof bounds Bψ1(x∗,x1)B_{\psi_1}(x^*,x_1)Bψ1​​(x∗,x1​) by 12∥x∗−x1∥∞2⟨1,s1⟩\frac12\|x^*-x_1\|_\infty^2\langle\mathbf 1,s_1\rangle21​∥x∗−x1​∥∞2​⟨1,s1​⟩, which drops δ2∥x∗−x1∥22\frac\delta2\|x^*-x_1\|_2^22δ​∥x∗−x1​∥22​. For large δ>0\delta>0δ>0 the printed bound is false. δ=0\delta=0δ=0 is the case the proof covers and the one the paper's corollaries use.
  2. φ\varphiφ is minimized over X\mathcal XX at x1=0x_1=0x1​=0. Propositions 2 and 3 rely on this ("x1=argmin⁡x∈Xφ(x)x_1=\operatorname{argmin}_{x\in\mathcal X}\varphi(x)x1​=argminx∈X​φ(x)", "w.l.o.g. φ(x1)=0\varphi(x_1)=0φ(x1​)=0"). It holds for φ=0\varphi=0φ=0, λ∥x∥1\lambda\|x\|_1λ∥x∥1​ and λ∥x∥22\lambda\|x\|_2^2λ∥x∥22​. Without it the bounds fail.

Proposition 2 is stated with strictly positive weights. The primal-dual part of the goal with δ=0\delta=0δ=0 is a degenerate case in which every gtg_tgt​ vanishes.

A run predicate without the subgradient link gt∈∂ft(xt)g_t\in\partial f_t(x_t)gt​∈∂ft​(xt​), an argmin without xt+1∈Xx_{t+1}\in\mathcal Xxt+1​∈X, a comparator outside X\mathcal XX, or sts_tst​ computed without round ttt would each make the statement trivial or different. The definitions rule all of these out.

Needed infrastructure: elementary convex analysis on Rd\mathbb R^dRd (first-order optimality of a convex objective over a convex set), finite sums and square roots. Nothing beyond Mathlib is required. The two regret templates (Propositions 2 and 3) and inequality (24) are reusable beyond this mission. Proofs of any milestone, and of the full-matrix analogues, are welcome.

Selected references

  • J. Duchi, E. Hazan, Y. Singer, Adaptive Subgradient Methods for Online Learning and Stochastic Optimization, Journal of Machine Learning Research 12 (2011) 2121–2159. https://jmlr.org/papers/v12/duchi11a.html
  • H. B. McMahan, M. Streeter, Adaptive Bound Optimization for Online Convex Optimization, COLT 2010. https://arxiv.org/abs/1002.4908
  • P. Auer, C. Gentile, Adaptive and Self-Confident On-Line Learning Algorithms, COLT 2000; J. Comput. System Sci. 64 (2002) 48–75. https://doi.org/10.1006/jcss.2001.1795
  • J. Duchi, S. Shalev-Shwartz, Y. Singer, A. Tewari, Composite Objective Mirror Descent, COLT 2010.
  • L. Xiao, Dual Averaging Methods for Regularized Stochastic Learning and Online Optimization, Journal of Machine Learning Research 11 (2010) 2543–2596. https://jmlr.org/papers/v11/xiao10a.html
  • Y. Nesterov, Primal-dual subgradient methods for convex problems, Mathematical Programming 120 (2009) 221–259. https://doi.org/10.1007/s10107-007-0149-x
  • M. Zinkevich, Online Convex Programming and Generalized Infinitesimal Gradient Ascent, ICML 2003.
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Convex OptimizationOptimization·Captain: mikedeng1

Convex Optimization: Algorithms and Complexity V: Projected Gradient Descent on a Convex β-Smooth Function Has Rate (3β‖x₁ − x*‖² + f(x₁) − f(x*))/tTextbook

Motivation

Many optimization problems in statistics, machine learning and operations research ask to minimize a smooth convex function over a simple convex set: a box, a ball, a simplex, a cone of positive semidefinite matrices. Projected gradient descent is the most basic first-order method for such problems. It takes a gradient step and then returns to the feasible set by Euclidean projection. Its cost per iteration is one gradient evaluation and one projection, independent of the dimension apart from the cost of manipulating vectors, which is why it remains a workhorse in very high dimension.

This mission is the fifth in a series formalizing S. Bubeck, Convex Optimization: Algorithms and Complexity (Foundations and Trends in Machine Learning 8(3–4), 2015; arXiv:1405.4980). Chapter 3 of the monograph treats "dimension-free" convex optimization. Its §3.2 first proves that gradient descent on a convex β\betaβ-smooth function on Rn\mathbb R^nRn has rate O(β∥x1−x∗∥2/t)O(\beta\|x_1-x^*\|^2/t)O(β∥x1​−x∗∥2/t) (Theorem 3.3, the subject of mission IV), and then turns to the constrained case, which is the subject of this mission. The analysis follows Nesterov's Introductory Lectures on Convex Optimization (2004).

Setting

Write x⊤yx^\top yx⊤y for the Euclidean inner product on Rn\mathbb R^nRn and ∥⋅∥\|\cdot\|∥⋅∥ for the Euclidean norm. The constraint set X⊆Rn\mathcal X\subseteq\mathbb R^nX⊆Rn is compact and convex; this is the standing assumption of Chapter 3. The Euclidean projection of y∈Rny\in\mathbb R^ny∈Rn onto X\mathcal XX is

ΠX(y)=argmin⁡z∈X∥y−z∥,\Pi_{\mathcal X}(y)=\operatorname*{argmin}_{z\in\mathcal X}\|y-z\| ,ΠX​(y)=z∈Xargmin​∥y−z∥,

which exists and is unique for a nonempty closed convex set.

A differentiable function f:Rn→Rf:\mathbb R^n\to\mathbb Rf:Rn→R is β\betaβ-smooth on X\mathcal XX if its gradient is β\betaβ-Lipschitz there:

∥∇f(x)−∇f(y)∥≤β∥x−y∥(x,y∈X).\|\nabla f(x)-\nabla f(y)\|\le\beta\|x-y\|\qquad(x,y\in\mathcal X).∥∇f(x)−∇f(y)∥≤β∥x−y∥(x,y∈X).

It is α\alphaα-strongly convex on X\mathcal XX if f(x)−f(y)≤∇f(x)⊤(x−y)−α2∥x−y∥2f(x)-f(y)\le\nabla f(x)^\top(x-y)-\frac\alpha2\|x-y\|^2f(x)−f(y)≤∇f(x)⊤(x−y)−2α​∥x−y∥2 for x,y∈Xx,y\in\mathcal Xx,y∈X. A minimizer x∗∈Xx^*\in\mathcal Xx∗∈X of fff over X\mathcal XX is assumed to exist, as everywhere in the monograph.

Projected gradient descent with step size η=1/β\eta=1/\betaη=1/β starts at x1∈Xx_1\in\mathcal Xx1​∈X and iterates

xt+1=ΠX(xt−1β∇f(xt))(t≥1).x_{t+1}=\Pi_{\mathcal X}\Bigl(x_t-\tfrac1\beta\nabla f(x_t)\Bigr)\qquad(t\ge1).xt+1​=ΠX​(xt​−β1​∇f(xt​))(t≥1).

For a point xxx with projected step x+=ΠX(x−1β∇f(x))x^+=\Pi_{\mathcal X}\bigl(x-\frac1\beta\nabla f(x)\bigr)x+=ΠX​(x−β1​∇f(x)), the gradient mapping is gX(x)=β(x−x+)g_{\mathcal X}(x)=\beta(x-x^+)gX​(x)=β(x−x+). Without constraints it equals ∇f(x)\nabla f(x)∇f(x).

Formalization targets

Goal: Theorem 3.7 (p. 270)

For fff convex and β\betaβ-smooth on X\mathcal XX, projected gradient descent with η=1/β\eta=1/\betaη=1/β satisfies, for every t≥1t\ge1t≥1,

f(xt)−f(x∗)≤3β∥x1−x∗∥2+f(x1)−f(x∗)t.f(x_t)-f(x^*)\le\frac{3\beta\|x_1-x^*\|^2+f(x_1)-f(x^*)}{t}.f(xt​)−f(x∗)≤t3β∥x1​−x∗∥2+f(x1​)−f(x∗)​.

Milestones

  1. Lemma 3.1 (p. 263): for x∈Xx\in\mathcal Xx∈X and y∈Rny\in\mathbb R^ny∈Rn, (ΠX(y)−x)⊤(ΠX(y)−y)≤0(\Pi_{\mathcal X}(y)-x)^\top(\Pi_{\mathcal X}(y)-y)\le0(ΠX​(y)−x)⊤(ΠX​(y)−y)≤0, and hence ∥ΠX(y)−x∥2+∥y−ΠX(y)∥2≤∥y−x∥2\|\Pi_{\mathcal X}(y)-x\|^2+\|y-\Pi_{\mathcal X}(y)\|^2\le\|y-x\|^2∥ΠX​(y)−x∥2+∥y−ΠX​(y)∥2≤∥y−x∥2. The first claim is an already proved platform theorem.
  2. (3.7) (p. 270): ∇f(x)⊤(x+−y)≤gX(x)⊤(x+−y)\nabla f(x)^\top(x^+-y)\le g_{\mathcal X}(x)^\top(x^+-y)∇f(x)⊤(x+−y)≤gX​(x)⊤(x+−y) for y∈Xy\in\mathcal Xy∈X.
  3. Lemma 3.6 (p. 270): for x,y∈Xx,y\in\mathcal Xx,y∈X,
f(x+)−f(y)≤gX(x)⊤(x−y)−12β∥gX(x)∥2.f(x^+)-f(y)\le g_{\mathcal X}(x)^\top(x-y)-\frac1{2\beta}\|g_{\mathcal X}(x)\|^2 .f(x+)−f(y)≤gX​(x)⊤(x−y)−2β1​∥gX​(x)∥2.
  1. Descent and gap (pp. 270–271): f(xs+1)−f(xs)≤−12β∥gX(xs)∥2f(x_{s+1})-f(x_s)\le-\frac1{2\beta}\|g_{\mathcal X}(x_s)\|^2f(xs+1​)−f(xs​)≤−2β1​∥gX​(xs​)∥2 and f(xs+1)−f(x∗)≤∥gX(xs)∥ ∥xs−x∗∥f(x_{s+1})-f(x^*)\le\|g_{\mathcal X}(x_s)\|\,\|x_s-x^*\|f(xs+1​)−f(x∗)≤∥gX​(xs​)∥∥xs​−x∗∥.
  2. Distances decrease (p. 271): ∥xs+1−x∗∥≤∥xs−x∗∥\|x_{s+1}-x^*\|\le\|x_s-x^*\|∥xs+1​−x∗∥≤∥xs​−x∗∥.
  3. Recursion and induction (p. 271): with δs=f(xs)−f(x∗)\delta_s=f(x_s)-f(x^*)δs​=f(xs​)−f(x∗), δs+1≤δs−12β∥x1−x∗∥2δs+12\delta_{s+1}\le\delta_s-\frac{1}{2\beta\|x_1-x^*\|^2}\delta_{s+1}^2δs+1​≤δs​−2β∥x1​−x∗∥21​δs+12​, and an induction on this recursion gives the bound of Theorem 3.7.

Companion: Theorem 3.10 (p. 278)

If fff is in addition α\alphaα-strongly convex on X\mathcal XX, with condition number κ=β/α\kappa=\beta/\alphaκ=β/α, then the strengthened Lemma 3.6, display (3.14), gives for every t≥0t\ge0t≥0

∥xt+1−x∗∥2≤exp⁡(−tκ)∥x1−x∗∥2.\|x_{t+1}-x^*\|^2\le\exp\Bigl(-\frac t\kappa\Bigr)\|x_1-x^*\|^2 .∥xt+1​−x∗∥2≤exp(−κt​)∥x1​−x∗∥2.

Significance

Theorem 3.7 shows that a constraint set with a cheap projection costs nothing in the order of convergence: the O(β∥x1−x∗∥2/t)O(\beta\|x_1-x^*\|^2/t)O(β∥x1​−x∗∥2/t) rate of unconstrained gradient descent survives, with a constant that depends on f(x1)−f(x∗)f(x_1)-f(x^*)f(x1​)−f(x∗). Lemma 3.6 is the reusable part. It isolates the gradient mapping as the quantity that measures progress under constraints, and the same inequality, with a strong-convexity term added, gives the linear rate of Theorem 3.10. It is also the starting point of the analysis of proximal gradient methods for composite objectives.

All of these results are classical and proved in the textbook. As far as a search of the platform shows, none of them is formalized there. A related platform theorem on constrained gradient descent for well-conditioned functions (from Hazan's Introduction to Online Convex Optimization) bounds function values with a different exponent. It is not Theorem 3.10, which bounds distances with rate exp⁡(−t/κ)\exp(-t/\kappa)exp(−t/κ). This mission adds machine-checked statements of the constrained analysis in Bubeck's formulation, with the book's constants. The definitions of β\betaβ-smoothness on a set, the gradient mapping and the run predicate can be reused by later missions on accelerated and proximal methods.

Difficulty

The obvious attempt is to copy the unconstrained proof, which rests on the descent inequality f(x−1β∇f(x))≤f(x)−12β∥∇f(x)∥2f(x-\frac1\beta\nabla f(x))\le f(x)-\frac1{2\beta}\|\nabla f(x)\|^2f(x−β1​∇f(x))≤f(x)−2β1​∥∇f(x)∥2. That inequality fails under constraints: the projection can cut the step short, so the decrease in fff is not controlled by ∥∇f(x)∥\|\nabla f(x)\|∥∇f(x)∥. At a boundary minimizer, for instance, ∇f(x∗)≠0\nabla f(x^*)\neq0∇f(x∗)=0 while no step makes progress. The proof has to replace the gradient by the gradient mapping throughout. It also has to show separately that the iterates do not move away from x∗x^*x∗, a fact that came for free without constraints from co-coercivity of the gradient (display (3.6)).

The closing induction is not routine either. The step from sss to s+1s+1s+1 works only from s=2s=2s=2 on, and the step from 111 to 222 needs a direct computation that uses the term f(x1)−f(x∗)f(x_1)-f(x^*)f(x1​)−f(x∗) in the numerator.

Formalization scope

  • Space. Rn\mathbb R^nRn is EuclideanSpace ℝ (Fin n), and x⊤yx^\top yx⊤y is ⟪x, y⟫_ℝ.
  • Constraint set and minimizer. Every statement assumes X\mathcal XX compact (IsCompact X) and convex (Convex ℝ X), the standing assumption of Chapter 3. Every statement about a run assumes a minimizer x∗∈Xx^*\in\mathcal Xx∗∈X with f(x∗)≤f(y)f(x^*)\le f(y)f(x∗)≤f(y) for all y∈Xy\in\mathcal Xy∈X, the monograph's standing assumption.
  • Gradient. The gradient is an explicit map g with HasGradientAt f (g x) x at every point of Rn\mathbb R^nRn, so ∇f(x)\nabla f(x)∇f(x) is defined at boundary points of X\mathcal XX. β\betaβ-smoothness on X\mathcal XX (IsBetaSmoothOn X f g β) asks this together with β≥0\beta\ge0β≥0 and the Lipschitz bound on X\mathcal XX only. It is not the quadratic upper bound (3.4), which is a consequence.
  • Strong convexity. It is the published predicate OnlineConvexOpt.ConvexBasics.StronglyConvexOn X f g α, which is display (3.13) with the same gradient map. Display (3.14) and Theorem 3.10 use α>0\alpha>0α>0.
  • Projection. The projection is the published relation IsMetricProjection X y p: p∈Xp\in\mathcal Xp∈X and ∥y−p∥≤∥y−z∥\|y-p\|\le\|y-z\|∥y−p∥≤∥y−z∥ for every z∈Xz\in\mathcal Xz∈X. No choice function is used.
  • Run. A run is IsProjGDRun X g β x: β>0\beta>0β>0, x1∈Xx_1\in\mathcal Xx1​∈X, and for every t≥1t\ge1t≥1 the iterate xt+1x_{t+1}xt+1​ is a projection of xt−β−1g(xt)x_t-\beta^{-1}g(x_t)xt​−β−1g(xt​). Indices start at 111 as in the book, and x 0 is unused.
  • Gradient mapping. gradMap β x xplus =β(x−x+)=\beta(x-x^+)=β(x−x+) takes the projected point of the run as an argument.
  • Positivity. β>0\beta>0β>0 is assumed, as the step size 1/β1/\beta1/β requires, and α>0\alpha>0α>0 for Theorem 3.10, as κ=β/α\kappa=\beta/\alphaκ=β/α requires.
  • The recursion. The recursion of milestone 6 is stated multiplied by 2β∥x1−x∗∥22\beta\|x_1-x^*\|^22β∥x1​−x∗∥2, so the case x1=x∗x_1=x^*x1​=x∗ needs no division.
  • No trivial readings. A run predicate with a free step size, or β\betaβ-smoothness replaced by the quadratic bound (3.4), would change the content. Both are excluded: the step is fixed to 1/β1/\beta1/β, and smoothness is the Lipschitz condition on the gradient.

Contributions are welcome on any milestone. Lemma 3.1 and (3.7) need only the variational characterization of the projection. Lemma 3.6 needs the quadratic upper bound for a function whose gradient is Lipschitz on a convex set, together with the first-order characterization of convexity relative to a set. Both are of independent use. The induction milestone is a self-contained statement about real sequences.

Selected references

  • S. Bubeck, Convex Optimization: Algorithms and Complexity, Foundations and Trends in Machine Learning 8(3–4):231–358, 2015. arXiv:1405.4980
  • Y. Nesterov, Introductory Lectures on Convex Optimization: A Basic Course, Kluwer Academic Publishers, 2004. doi:10.1007/978-1-4419-8853-9
  • E. Hazan, Introduction to Online Convex Optimization, 2nd ed., MIT Press, 2022. arXiv:1909.05207
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Convex OptimizationOptimization·Captain: mikedeng1

Convex Optimization: Algorithms and Complexity III: Projected Subgradient Descent with η = R/(L√t) Satisfies f(average) − f(x*) ≤ RL/√tTextbook

Motivation

Many convex optimization problems in machine learning and statistics have objectives that are convex but not differentiable: hinge losses, ℓ1\ell_1ℓ1​ penalties, maxima of finitely many affine functions, and the dual functions of Lagrangian relaxations. Methods that rely on gradients do not apply to them directly, while cutting-plane methods such as the ellipsoid method pay a price that grows with the dimension. The projected subgradient method replaces the gradient by an arbitrary subgradient and restores feasibility by a Euclidean projection. Its guarantee depends on the dimension only through two constants, a radius RRR and a Lipschitz constant LLL. This is the reason it, and its descendants (mirror descent, stochastic gradient descent, online gradient descent), are the standard tools for large-scale nonsmooth problems.

The rate analysed here goes back to the subgradient methods of Shor and Polyak in the 1960s and 1970s and to the lower bounds of Nemirovski and Yudin (1983). The book follows the presentation of Nesterov, Introductory Lectures on Convex Optimization (2004). The strongly convex variant with weights proportional to sss is from Lacoste-Julien, Schmidt and Bach (2012).

This mission is the third of a series formalizing S. Bubeck, Convex Optimization: Algorithms and Complexity (Foundations and Trends in Machine Learning, 2015), and covers the preamble of Chapter 3, Section 3.1 and Section 3.4.1.

Setting

Let Rn\mathbb R^nRn carry the Euclidean inner product x⊤yx^\top yx⊤y and norm ∥⋅∥\|\cdot\|∥⋅∥. Let X⊆Rn\mathcal X\subseteq\mathbb R^nX⊆Rn be compact and convex, and let fff be a convex function on X\mathcal XX with a minimizer x∗∈Xx^*\in\mathcal Xx∗∈X.

A vector ggg is a subgradient of fff at x∈Xx\in\mathcal Xx∈X if f(x)−f(y)≤g⊤(x−y)f(x)-f(y)\le g^\top(x-y)f(x)−f(y)≤g⊤(x−y) for every y∈Xy\in\mathcal Xy∈X. The set of subgradients at xxx is written ∂f(x)\partial f(x)∂f(x). The projection ΠX(y)\Pi_{\mathcal X}(y)ΠX​(y) of a point y∈Rny\in\mathbb R^ny∈Rn is the point of X\mathcal XX nearest to yyy.

Fix step sizes ηs>0\eta_s>0ηs​>0. Projected subgradient descent starts at some x1∈Xx_1\in\mathcal Xx1​∈X and iterates, for s≥1s\ge1s≥1,

ys+1=xs−ηsgs,gs∈∂f(xs),xs+1=ΠX(ys+1).y_{s+1}=x_s-\eta_s g_s,\quad g_s\in\partial f(x_s),\qquad x_{s+1}=\Pi_{\mathcal X}(y_{s+1}).ys+1​=xs​−ηs​gs​,gs​∈∂f(xs​),xs+1​=ΠX​(ys+1​).

Any subgradient may be chosen at each step. In Section 3.1 the step is constant, ηs=η\eta_s=\etaηs​=η. The set X\mathcal XX lies in the Euclidean ball of radius RRR centred at x1x_1x1​, and the subgradients have norm at most LLL.

A function fff is α\alphaα-strongly convex on X\mathcal XX if f(x)−f(y)≤g⊤(x−y)−α2∥x−y∥2f(x)-f(y)\le g^\top(x-y)-\frac{\alpha}{2}\|x-y\|^2f(x)−f(y)≤g⊤(x−y)−2α​∥x−y∥2 for all x,y∈Xx,y\in\mathcal Xx,y∈X and g∈∂f(x)g\in\partial f(x)g∈∂f(x).

Formalization targets

Goal: Theorem 3.2

For every horizon t≥1t\ge1t≥1, projected subgradient descent with the constant step η=R/(Lt)\eta=R/(L\sqrt t)η=R/(Lt​) satisfies

f(1t∑s=1txs)−f(x∗)≤RLt.f\Big(\frac1t\sum_{s=1}^{t}x_s\Big)-f(x^*)\le\frac{RL}{\sqrt t}.f(t1​s=1∑t​xs​)−f(x∗)≤t​RL​.

Milestones

  1. Lemma 3.1. For x∈Xx\in\mathcal Xx∈X and y∈Rny\in\mathbb R^ny∈Rn: (ΠX(y)−x)⊤(ΠX(y)−y)≤0(\Pi_{\mathcal X}(y)-x)^\top(\Pi_{\mathcal X}(y)-y)\le0(ΠX​(y)−x)⊤(ΠX​(y)−y)≤0, already on the platform as a published theorem. The mission also states its consequence
∥ΠX(y)−x∥2+∥y−ΠX(y)∥2≤∥y−x∥2.\|\Pi_{\mathcal X}(y)-x\|^2+\|y-\Pi_{\mathcal X}(y)\|^2\le\|y-x\|^2 .∥ΠX​(y)−x∥2+∥y−ΠX​(y)∥2≤∥y−x∥2.
  1. The per-step inequality in the proof of Theorem 3.2:
f(xs)−f(x∗)≤12η(∥xs−x∗∥2−∥ys+1−x∗∥2)+η2∥gs∥2.f(x_s)-f(x^*)\le\frac1{2\eta}\big(\|x_s-x^*\|^2-\|y_{s+1}-x^*\|^2\big)+\frac\eta2\|g_s\|^2 .f(xs​)−f(x∗)≤2η1​(∥xs​−x∗∥2−∥ys+1​−x∗∥2)+2η​∥gs​∥2.
  1. The summed inequality for any constant step η>0\eta>0η>0:
∑s=1t(f(xs)−f(x∗))≤R22η+ηL2t2.\sum_{s=1}^{t}\big(f(x_s)-f(x^*)\big)\le\frac{R^2}{2\eta}+\frac{\eta L^2t}{2}.s=1∑t​(f(xs​)−f(x∗))≤2ηR2​+2ηL2t​.

Companion: Theorem 3.9

If fff is α\alphaα-strongly convex and its subgradients are bounded by LLL, then with ηs=2/(α(s+1))\eta_s=2/(\alpha(s+1))ηs​=2/(α(s+1)),

f(∑s=1t2st(t+1)xs)−f(x∗)≤2L2α(t+1).f\Big(\sum_{s=1}^{t}\frac{2s}{t(t+1)}x_s\Big)-f(x^*)\le\frac{2L^2}{\alpha(t+1)}.f(s=1∑t​t(t+1)2s​xs​)−f(x∗)≤α(t+1)2L2​.

Significance

Theorem 3.2 gives an oracle complexity of O(R2L2/ε2)O(R^2L^2/\varepsilon^2)O(R2L2/ε2) for reaching an ε\varepsilonε-optimal point, independent of the ambient dimension. Section 3.5 of the book shows this rate is unimprovable for black-box first-order methods once the dimension is large. Theorem 3.9 shows how strong convexity improves the rate to O(1/t)O(1/t)O(1/t), with the averaging weights changed from uniform to linear. These two bounds are the reference points against which the rest of Chapter 3 and Chapters 4 to 6 (smooth, accelerated, mirror, stochastic methods) are measured.

The results are classical and their proofs are short. Formalizing them produces a reusable, machine-checked account of the basic projected first-order step: the projection inequality, the one-step distance recursion, and the telescoping argument with Jensen's inequality for averaged iterates. To our knowledge, no machine-checked proof of the averaged-iterate bound for the projected subgradient method exists in Mathlib. Related platform items cover other algorithms or other averaging schemes.

Difficulty

The arithmetic is elementary, and the obvious argument works. The care is in the bookkeeping. The projection must be shown not to increase the distance to x∗x^*x∗, which needs convexity of X\mathcal XX and the variational characterization of the nearest point. The sum must telescope with a horizon-dependent constant step. Jensen's inequality must be applied to a finite convex combination of points of X\mathcal XX, which requires showing that the average lies in X\mathcal XX. In Theorem 3.9 the step sizes and the averaging weights are coupled, so neither can be changed independently. In Lean, the iterates are indexed from 111 with natural-number horizons, and the bounds involve t\sqrt tt​, so these casts need care.

Formalization scope

Rn\mathbb R^nRn is EuclideanSpace ℝ (Fin n). The iterates are sequences ℕ → EuclideanSpace ℝ (Fin n) with the first iterate at index 111. The projection is the published relation OnlineConvexOpt.FirstOrder.IsMetricProjection (xs+1∈Xx_{s+1}\in\mathcal Xxs+1​∈X is a nearest point to ys+1y_{s+1}ys+1​). Subgradients are taken relative to X\mathcal XX (Definition 1.2). A run is a predicate on the steps s=1,…,ts=1,\dots,ts=1,…,t, and every theorem holds for all runs, that is, for every choice of subgradients. Compactness and convexity of X\mathcal XX, convexity of fff on X\mathcal XX and the existence of the minimizer x∗x^*x∗ are the book's standing assumptions and appear as hypotheses. R>0R>0R>0, L>0L>0L>0 and α>0\alpha>0α>0 are explicit, because Lean's division by zero would otherwise turn the step size into a junk value.

The book assumes ∥g∥≤L\|g\|\le L∥g∥≤L for every subgradient at every point of X\mathcal XX. With subgradients relative to a compact X\mathcal XX, that assumption can never hold at a boundary point, since every outward normal can be added to a subgradient. Stated that way the theorems would be vacuous. The mission therefore assumes the bound only for the subgradients g1,…,gtg_1,\dots,g_tg1​,…,gt​ that the run uses. This is a weaker hypothesis and gives a stronger, non-vacuous statement. Bounding only these subgradients is a deliberate choice, not a trivialization: the bound still constrains every quantity the conclusion depends on.

Contributions welcome: proofs of the four inequalities and of the two rates, and a reusable lemma that the convex combination of finitely many points of a convex set lies in the set, together with Jensen's inequality in the form used here.

Selected references

  • S. Bubeck, Convex Optimization: Algorithms and Complexity, Foundations and Trends in Machine Learning 8(3–4):231–358, 2015. https://arxiv.org/abs/1405.4980
  • Y. Nesterov, Introductory Lectures on Convex Optimization: A Basic Course, Kluwer, 2004. https://doi.org/10.1007/978-1-4419-8853-9
  • A. Nemirovski and D. Yudin, Problem Complexity and Method Efficiency in Optimization, Wiley, 1983.
  • S. Lacoste-Julien, M. Schmidt and F. Bach, A simpler approach to obtaining an O(1/t) convergence rate for the projected stochastic subgradient method, 2012. https://arxiv.org/abs/1212.2002
  • N. Z. Shor, Minimization Methods for Non-Differentiable Functions, Springer, 1985. https://doi.org/10.1007/978-3-642-82118-9
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ProbabilityStatistics·Captain: mikedeng1

Local Rademacher Complexities II: Local Rademacher Averages of the Classification Loss Class Are Bounded by Weighted Empirical Risk Minimization (Theorem 6.3)Research Paper

Motivation

Local Rademacher averages measure the complexity of a learning problem only near the functions that matter, such as those with small empirical error, rather than over the whole function class. Bartlett, Bousquet and Mendelson (Local Rademacher complexities, Ann. Statist. 33 (2005)) show that error bounds for empirical risk minimization are governed by the fixed point of a sub-root upper bound on such local averages, and that these bounds give fast rates (1/n1/n1/n rather than 1/n1/\sqrt n1/n​) under variance conditions. A bound is only useful in practice if it can be computed from the data. For classification with the discrete loss, the paper's Corollary 6.2 states the bound in terms of a localized empirical Rademacher average ψ^n(r)\hat\psi_n(r)ψ^​n​(r). That average is a supremum over a constrained subclass, and it is not obvious how to evaluate it.

Theorem 6.3 of the paper answers this. An upper bound on ψ^n(r)\hat\psi_n(r)ψ^​n​(r) can be computed by any algorithm that minimizes a weighted empirical classification error. A similar reduction was known for the global Rademacher average of a classification class: Bartlett, Boucheron and Lugosi (Model selection and error estimation, Machine Learning 48 (2002)) observed that the empirical Rademacher average equals one half minus an expected empirical risk minimum with random labels, and Lemma 6.4 of the paper is adapted from their argument. Theorem 6.3 shows that localization and the use of star-hulls keep this reduction intact.

Setting

Fix inputs X1,…,XnX_1,\dots,X_nX1​,…,Xn​ in a set X\mathcal XX (n≥1n\ge1n≥1) and labels Y1,…,Yn∈{−1,1}Y_1,\dots,Y_n\in\{-1,1\}Y1​,…,Yn​∈{−1,1}. Everything below is deterministic given this sample. A classifier is a function f:X→{−1,1}f:\mathcal X\to\{-1,1\}f:X→{−1,1}, and F\mathcal FF is a class of classifiers. The discrete loss is ℓ(y,y′)=1[y≠y′]\ell(y,y')=\mathbf 1[y\ne y']ℓ(y,y′)=1[y=y′]. For a vector z∈Rnz\in\mathbb R^nz∈Rn write

Pnℓ(f(X),z)=1n∑i=1nℓ(f(Xi),zi),Pnℓf=Pnℓ(f(X),Y),P_n\ell(f(X),z)=\frac1n\sum_{i=1}^n\ell(f(X_i),z_i),\qquad P_n\ell_f=P_n\ell(f(X),Y),Pn​ℓ(f(X),z)=n1​i=1∑n​ℓ(f(Xi​),zi​),Pn​ℓf​=Pn​ℓ(f(X),Y),

so PnℓfP_n\ell_fPn​ℓf​ is the empirical risk of fff.

A sign vector σ∈{−1,1}n\sigma\in\{-1,1\}^nσ∈{−1,1}n plays the role of Rademacher signs, and Eσ\mathbb E_\sigmaEσ​ is the average over all 2n2^n2n sign vectors. The empirical Rademacher average of the loss functions of the classifiers with empirical risk at most bbb is

EσRn{ℓf:f∈F, Pnℓf≤b}=1n Eσsup⁡f∈F, Pnℓf≤b ∑i=1nσi ℓ(f(Xi),Yi).\mathbb E_\sigma R_n\{\ell_f : f\in\mathcal F,\ P_n\ell_f\le b\}=\frac1n\,\mathbb E_\sigma\sup_{f\in\mathcal F,\ P_n\ell_f\le b}\ \sum_{i=1}^n\sigma_i\,\ell(f(X_i),Y_i).Eσ​Rn​{ℓf​:f∈F, Pn​ℓf​≤b}=n1​Eσ​f∈F, Pn​ℓf​≤bsup​ i=1∑n​σi​ℓ(f(Xi​),Yi​).

For c≥0c\ge0c≥0, x>0x>0x>0 and 0<r≤1/20<r\le1/20<r≤1/2, the empirical local Rademacher complexity of the classification loss class is

ψ^n(r)=csup⁡α∈[2r,1]α EσRn{ℓf:f∈F, Pnℓf≤2r/α2}+26xn.\hat\psi_n(r)=c\sup_{\alpha\in[\sqrt{2r},1]}\alpha\,\mathbb E_\sigma R_n\{\ell_f : f\in\mathcal F,\ P_n\ell_f\le 2r/\alpha^2\}+\frac{26x}{n}.ψ^​n​(r)=cα∈[2r​,1]sup​αEσ​Rn​{ℓf​:f∈F, Pn​ℓf​≤2r/α2}+n26x​.

In Corollary 6.2, c=20c=20c=20. The parameter α\alphaα comes from the star-hull of the loss class: rescaling a loss function by α\alphaα turns the constraint Pn(αℓf)2≤2rP_n(\alpha\ell_f)^2\le 2rPn​(αℓf​)2≤2r into Pnℓf≤2r/α2P_n\ell_f\le 2r/\alpha^2Pn​ℓf​≤2r/α2.

For a sign vector σ\sigmaσ and a multiplier μ≥0\mu\ge0μ≥0, the weighted empirical risk minimum is

J(μ)=min⁡f∈F1n∑i=1n∣σi+μYi∣ ℓ(f(Xi),sign⁡(σi+μYi)).J(\mu)=\min_{f\in\mathcal F}\frac1n\sum_{i=1}^n|\sigma_i+\mu Y_i|\,\ell\big(f(X_i),\operatorname{sign}(\sigma_i+\mu Y_i)\big).J(μ)=f∈Fmin​n1​i=1∑n​∣σi​+μYi​∣ℓ(f(Xi​),sign(σi​+μYi​)).

It is the smallest weighted training error when the labels are corrupted to sign⁡(σi+μYi)\operatorname{sign}(\sigma_i+\mu Y_i)sign(σi​+μYi​) and example iii has weight ∣σi+μYi∣|\sigma_i+\mu Y_i|∣σi​+μYi​∣.

Formalization targets

Goal: Theorem 6.3

If some f∈Ff\in\mathcal Ff∈F has Pnℓf≤2rP_n\ell_f\le 2rPn​ℓf​≤2r, then

ψ^n(r)≤csup⁡α∈[2r,1]α Eσmin⁡μ≥0((2rα2−12)μ+12n∑i=1n∣σi+μYi∣−J(μ))+26xn.\hat\psi_n(r)\le c\sup_{\alpha\in[\sqrt{2r},1]}\alpha\,\mathbb E_\sigma\min_{\mu\ge0}\Big(\Big(\frac{2r}{\alpha^2}-\frac12\Big)\mu+\frac1{2n}\sum_{i=1}^n|\sigma_i+\mu Y_i|-J(\mu)\Big)+\frac{26x}{n}.ψ^​n​(r)≤cα∈[2r​,1]sup​αEσ​μ≥0min​((α22r​−21​)μ+2n1​i=1∑n​∣σi​+μYi​∣−J(μ))+n26x​.

The multiplier ccc is kept general, and the term 26x/n26x/n26x/n appears on both sides as printed.

Milestones

  1. Lemma 6.4. For every b∈[0,1]b\in[0,1]b∈[0,1] with a feasible classifier,
EσRn{ℓf:f∈F, Pnℓf≤b}=12−Eσmin⁡{Pnℓ(f(X),σ):f∈F, Pnℓ(f(X),Y)≤b}.\mathbb E_\sigma R_n\{\ell_f : f\in\mathcal F,\ P_n\ell_f\le b\}=\frac12-\mathbb E_\sigma\min\{P_n\ell(f(X),\sigma) : f\in\mathcal F,\ P_n\ell(f(X),Y)\le b\}.Eσ​Rn​{ℓf​:f∈F, Pn​ℓf​≤b}=21​−Eσ​min{Pn​ℓ(f(X),σ):f∈F, Pn​ℓ(f(X),Y)≤b}.
  1. Weak duality (proof of Theorem 6.3). With L(f,μ)=Pnℓ(f(X),σ)+μ(Pnℓ(f(X),Y)−2r/α2)L(f,\mu)=P_n\ell(f(X),\sigma)+\mu(P_n\ell(f(X),Y)-2r/\alpha^2)L(f,μ)=Pn​ℓ(f(X),σ)+μ(Pn​ℓ(f(X),Y)−2r/α2) and g(μ)=min⁡f∈FL(f,μ)g(\mu)=\min_{f\in\mathcal F}L(f,\mu)g(μ)=minf∈F​L(f,μ), for every μ≥0\mu\ge0μ≥0,
min⁡{Pnℓ(f(X),σ):f∈F, Pnℓ(f(X),Y)≤2r/α2}≥g(μ).\min\{P_n\ell(f(X),\sigma) : f\in\mathcal F,\ P_n\ell(f(X),Y)\le 2r/\alpha^2\}\ge g(\mu).min{Pn​ℓ(f(X),σ):f∈F, Pn​ℓ(f(X),Y)≤2r/α2}≥g(μ).
  1. The identity for g(μ)g(\mu)g(μ) (proof of Theorem 6.3, corrected).
g(μ)=J(μ)−12n∑i=1n∣σi+μYi∣+1+μ2−μ2rα2.g(\mu)=J(\mu)-\frac1{2n}\sum_{i=1}^n|\sigma_i+\mu Y_i|+\frac{1+\mu}2-\mu\frac{2r}{\alpha^2}.g(μ)=J(μ)−2n1​i=1∑n​∣σi​+μYi​∣+21+μ​−μα22r​.

Significance

The theorem turns a quantity defined by a supremum over a data-dependent subclass into one computable by a standard learning primitive. For each sign vector and each multiplier μ\muμ, J(μ)J(\mu)J(μ) is the value of a weighted classification problem, which any weighted empirical risk minimizer solves. The expectation over signs can be estimated by repeated sampling. The paper notes that JJJ is Lipschitz in μ\muμ, so a finite grid of μ\muμ values suffices, and that a sub-root upper bound on ψ^n\hat\psi_nψ^​n​ can then be read off. Combined with Corollary 6.2, this yields error bounds for empirical risk minimization in classification that are computable from the training data and that localize: they depend only on the classifiers with small empirical error.

The result is proved in the paper. It has not been formalized; as far as a search of the Prove2Me catalog shows, neither the classification loss class nor J(μ)J(\mu)J(μ) exists as a formal object. This mission produces machine-checked statements of the theorem and of its three proof steps. These cover the exact identity between Rademacher averages of the discrete loss class and random-label empirical risk minimization, and a Lagrangian duality bound for constrained empirical risk minimization.

Difficulty

The obvious route is to apply Lemma 6.4 and then exchange the constrained minimum for a Lagrangian. Each step has a point where a careless argument fails.

  • Lemma 6.4 needs a change of variables on sign vectors (σi↦−Yiσi\sigma_i\mapsto-Y_i\sigma_iσi​↦−Yi​σi​) that preserves the uniform average. It also needs the identity ℓ(y,y′)=∣y−y′∣/2\ell(y,y')=|y-y'|/2ℓ(y,y′)=∣y−y′∣/2 on {±1}\{\pm1\}{±1}, which fails off {±1}\{\pm1\}{±1}.
  • The Lagrangian step gives only weak duality. The bound is an inequality, and attempts to prove equality in Theorem 6.3 fail in general.
  • The identity for g(μ)g(\mu)g(μ) rests on ℓ(y,y^)=(1−yy^)/2\ell(y,\hat y)=(1-y\hat y)/2ℓ(y,y^​)=(1−yy^​)/2. This holds only for ±1\pm1±1 arguments, while sign⁡(σi+μYi)\operatorname{sign}(\sigma_i+\mu Y_i)sign(σi​+μYi​) is 000 when μ=1\mu=1μ=1 and σi=−Yi\sigma_i=-Y_iσi​=−Yi​. Those terms carry weight zero, and the bookkeeping has to show this.
  • Passing the per-α\alphaα, per-σ\sigmaσ inequalities through the outer supremum and the average requires every supremum and minimum to be over a nonempty, bounded set. This is where the feasibility hypothesis is used.

Formalization scope

  • Representation. Inputs are xs : Fin n → X for an arbitrary type X. Labels and signs are real vectors Fin n → ℝ. Classifiers are functions X → ℝ with values in {±1}\{\pm1\}{±1}, a class is a Set (X → ℝ), and the discrete loss is defined on all real pairs. Sign vectors are indexed by Fin n → Bool through the published UnderstandingML.signVec. Every Eσ\mathbb E_\sigmaEσ​, on both sides of every statement, is the finite average over these 2n2^n2n vectors; no probability measure is used. The empirical Rademacher average is the published UnderstandingML.rademacher applied to the set of loss vectors.
  • Suprema and minima. Every supremum and minimum is Lean's real ⨆/⨅ over a subtype. The hypotheses make each index set nonempty and each family bounded, so these are true suprema and minima. The convention Real.sign 0 = 0 is used where the paper's sign is undefined; it affects only weight-zero terms.
  • Added hypotheses. Each of these is implicit on the page:
    • n≥1n\ge1n≥1;
    • c≥0c\ge0c≥0 (for c<0c<0c<0 the inequality reverses);
    • 0<r≤1/20<r\le1/20<r≤1/2 (otherwise the range of α\alphaα is empty);
    • x>0x>0x>0 (Corollary 6.2's "fix x>0x>0x>0");
    • a classifier with Pnℓf≤2rP_n\ell_f\le 2rPn​ℓf​≤2r in the goal, and a feasible classifier in Lemma 6.4 and in the weak duality step (the page's minima presuppose one);
    • a nonempty F\mathcal FF in the g(μ)g(\mu)g(μ) identity.
  • Corrections of the print. The last display of the proof on p. 30 ends each line with −2r/α2-2r/\alpha^2−2r/α2. From the page's own definition g(μ)=min⁡fL(f,μ)g(\mu)=\min_f L(f,\mu)g(μ)=minf​L(f,μ), the constant is −μ 2r/α2-\mu\,2r/\alpha^2−μ2r/α2, which is the form Theorem 6.3's term (2r/α2−1/2)μ(2r/\alpha^2-1/2)\mu(2r/α2−1/2)μ requires. The milestone states the corrected identity.
  • No trivialization. Without the feasibility hypothesis, Lean would evaluate the empty-class Rademacher average and the unbounded μ\muμ-minimum to the junk value 000, and the goal would compare junk values. The feasibility hypothesis rules this out. The goal is the inequality between the two expressions for ψ^n\hat\psi_nψ^​n​ as printed; it is not restated through g(μ)g(\mu)g(μ), L(f,μ)L(f,\mu)L(f,μ) or Lemma 6.4.
  • Contributions welcome. A reusable lemma that the uniform average over {±1}n\{\pm1\}^n{±1}n is invariant under coordinatewise sign flips would serve beyond this mission, as would general facts about real infima over finite-valued families. Proofs of the three milestones, and of the goal from them, are the main targets.

Selected references

  • P. L. Bartlett, O. Bousquet, S. Mendelson, Local Rademacher complexities, Annals of Statistics 33(4), 1497–1537, 2005. arXiv:math/0508275v1 (cited version): https://arxiv.org/abs/math/0508275, DOI https://doi.org/10.1214/009053605000000282 — §6.2, Corollary 6.2 and Theorem 6.3 (pp. 28–29), Lemma 6.4 (p. 29), proof of Theorem 6.3 (p. 30).
  • P. L. Bartlett, S. Boucheron, G. Lugosi, Model selection and error estimation, Machine Learning 48, 85–113, 2002. https://doi.org/10.1023/A:1013999503812
  • P. L. Bartlett, S. Mendelson, Rademacher and Gaussian complexities: risk bounds and structural results, Journal of Machine Learning Research 3, 463–482, 2002. https://www.jmlr.org/papers/v3/bartlett02a.html
  • S. Shalev-Shwartz, S. Ben-David, Understanding Machine Learning: From Theory to Algorithms, Cambridge University Press, 2014, Chapter 26 (the Rademacher complexity reused here). https://doi.org/10.1017/CBO9781107298019
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Operations ResearchStatistics·Captain: mikedeng1

The Big Data Newsvendor: Practical Insights from Machine Learning: The L2-Regularized Feature-Based Newsvendor Rule Generalizes with a Bound Free of the Number of FeaturesResearch Paper

Motivation

The newsvendor problem is the basic model of inventory under uncertain demand. A decision maker orders qqq units before demand DDD is observed, and pays a unit backordering cost bbb for each unit of unmet demand and a unit holding cost hhh for each unit left over. When the demand distribution is known, the optimal order is a quantile of it. In practice the distribution is unknown and the decision maker holds historical data, often including features: observable covariates such as the day of the week, the weather or a recent sales trend, recorded alongside each past demand.

Rudin and Vahn (MIT Sloan Working Paper 5036-13, version of February 6, 2014, from MIT DSpace; published as Ban and Rudin, Operations Research 67(1), 2019, doi:10.1287/opre.2018.1757) propose to learn the order quantity directly as a linear function of the features, by minimizing the empirical newsvendor cost over the training data, with or without a regularization penalty. Their question is the one any data-driven decision rule must answer: how much worse can the learned rule do on new data than it did on the data it was fitted to? When the number of features ppp is comparable to the sample size nnn ("big data"), a bound that grows with ppp says nothing, and the paper's Theorem 2 gives one for the regularized rule that does not depend on ppp.

This mission formalizes that bound, Theorem 2 of the working paper, together with the results its proof is assembled from. All page and result numbers refer to the 2014 working paper, not to the published article.

Setting

Cost. For an order qqq and a demand ddd, the newsvendor cost is

C(q;d)=b (d−q)++h (q−d)+,b,h>0.C(q;d)=b\,(d-q)^+ + h\,(q-d)^+ ,\qquad b,h>0 .C(q;d)=b(d−q)++h(q−d)+,b,h>0.

Write b∨h=max⁡(b,h)b\vee h=\max(b,h)b∨h=max(b,h).

Data. A data point is a pair z=(x,d)z=(x,d)z=(x,d) of a feature vector x∈Rpx\in\mathbb R^px∈Rp and a demand d∈Rd\in\mathbb Rd∈R. Features lie in a domain X\mathcal XX inside the ball ∥x∥22≤Xmax⁡2\|x\|_2^2\le X_{\max}^2∥x∥22​≤Xmax2​, and demands lie in D=[0,Dˉ]\mathcal D=[0,\bar D]D=[0,Dˉ]. A sample Sn={(xi,di)}i=1nS_n=\{(x_i,d_i)\}_{i=1}^nSn​={(xi​,di​)}i=1n​ consists of nnn independent draws from an unknown probability distribution μ\muμ concentrated on X×D\mathcal X\times\mathcal DX×D.

Rules and risks. A vector q∈Rpq\in\mathbb R^pq∈Rp defines the linear decision rule q(x)=q⊤xq(x)=q^\top xq(x)=q⊤x. Its true risk and empirical risk are

Rtrue(q)=E(x,d)∼μ[C(q⊤x;d)],R^(q;Sn)=1n∑i=1nC(q⊤xi;di).R_{true}(q)=\mathbb E_{(x,d)\sim\mu}\bigl[C(q^\top x;d)\bigr],\qquad \hat R(q;S_n)=\frac1n\sum_{i=1}^n C(q^\top x_i;d_i).Rtrue​(q)=E(x,d)∼μ​[C(q⊤x;d)],R^(q;Sn​)=n1​i=1∑n​C(q⊤xi​;di​).

The regularized algorithm (NV-reg). For a parameter λ>0\lambda>0λ>0, the rule q^=q^(Sn)\hat q=\hat q(S_n)q^​=q^​(Sn​) minimizes

R^(q;Sn)+λ∥q∥22over q∈Rp.\hat R(q;S_n)+\lambda\|q\|_2^2\qquad\text{over } q\in\mathbb R^p .R^(q;Sn​)+λ∥q∥22​over q∈Rp.

The objective is strictly convex, so the minimizer is unique. Following Appendix B of the paper, the rules the algorithm outputs on samples from X×D\mathcal X\times\mathcal DX×D are assumed to map X\mathcal XX into D\mathcal DD (the paper's Q⊂DX\mathcal Q\subset\mathcal D^{\mathcal X}Q⊂DX): the learned order quantity is never negative and never exceeds the demand cap.

Uniform stability. An algorithm is uniformly stable with parameter αn\alpha_nαn​ if removing one observation from any sample changes the loss of its output at any test point by at most αn\alpha_nαn​ (Bousquet and Elisseeff's Definition 6, the paper's Definition 1).

Formalization targets

Goal: Theorem 2 (p. 9)

For every δ∈(0,1)\delta\in(0,1)δ∈(0,1) and n≥1n\ge1n≥1, with probability at least 1−δ1-\delta1−δ over SnS_nSn​,

∣Rtrue(q^)−R^(q^;Sn)∣≤(b∨h)2Xmax⁡2nλ+(2(b∨h)2Xmax⁡2λ+(b∨h)Dˉ)ln⁡(2/δ)2n.|R_{true}(\hat q)-\hat R(\hat q;S_n)|\le\frac{(b\vee h)^2X_{\max}^2}{n\lambda}+\Bigl(\frac{2(b\vee h)^2X_{\max}^2}{\lambda}+(b\vee h)\bar D\Bigr)\sqrt{\frac{\ln(2/\delta)}{2n}} .∣Rtrue​(q^​)−R^(q^​;Sn​)∣≤nλ(b∨h)2Xmax2​​+(λ2(b∨h)2Xmax2​​+(b∨h)Dˉ)2nln(2/δ)​​.

The dimension ppp appears nowhere in the bound.

Milestones, in the order of the proof (p. 32)

  1. Lemma 5 (p. 28). For q,d∈[0,Dˉ]q,d\in[0,\bar D]q,d∈[0,Dˉ], ∣C(q;d)∣≤(b∨h)Dˉ|C(q;d)|\le(b\vee h)\bar D∣C(q;d)∣≤(b∨h)Dˉ, and the bound is attained.
  2. Display (31) (p. 31). CCC is convex in its first argument and (b∨h)(b\vee h)(b∨h)-Lipschitz in it, i.e. (b∨h)(b\vee h)(b∨h)-admissible in the sense of Definition 2.
  3. Theorem 5 (p. 31), Bousquet and Elisseeff's Theorem 22: regularization in a reproducing kernel Hilbert space with a σ\sigmaσ-admissible loss and kernel bound κ2\kappa^2κ2 has uniform stability σ2κ2/(2λn)\sigma^2\kappa^2/(2\lambda n)σ2κ2/(2λn). This is an existing platform statement, referenced rather than restated.
  4. Theorem 4 (p. 30). (NV-reg) is uniformly stable with parameter αnr=(b∨h)2Xmax⁡2/(2nλ)\alpha_n^r=(b\vee h)^2X_{\max}^2/(2n\lambda)αnr​=(b∨h)2Xmax2​/(2nλ).
  5. Theorem 6 (p. 31). Any algorithm with uniform stability αn\alpha_nαn​ and loss in [0,M][0,M][0,M] satisfies, with probability at least 1−δ1-\delta1−δ,
∣Rtrue(A,Sn)−R^(A,Sn)∣≤2αn+(4nαn+M)ln⁡(2/δ)2n.|R_{true}(A,S_n)-\hat R(A,S_n)|\le2\alpha_n+(4n\alpha_n+M)\sqrt{\frac{\ln(2/\delta)}{2n}} .∣Rtrue​(A,Sn​)−R^(A,Sn​)∣≤2αn​+(4nαn​+M)2nln(2/δ)​​.

Significance

The result. Theorem 2 bounds the generalization gap of the regularized feature-based newsvendor rule at rate O(1/n)O(1/\sqrt n)O(1/n​) with constants that depend on the costs, the demand cap, the feature radius and λ\lambdaλ, but not on the number of features. It gives a theoretical basis for regularizing when p/np/np/n is not small, and it indicates how to scale λ\lambdaλ with the feature radius. The companion bound for the unregularized rule (Theorem 1) grows linearly in ppp.

Formalizing it. The theorem is proved in the paper, but its proof is short and relies on cited results: Theorem 5 and Theorem 6 are stated with references to Bousquet and Elisseeff and no proof of their own, and the paper's Theorem 6 is a two-sided variant of Bousquet and Elisseeff's Theorem 12 that is only sketched. None of these results has a machine-checked proof. A formalization produces a checked two-sided stability-to-generalization theorem for general algorithms (reusable for any stable learner), a checked stability bound for a regularized piecewise-linear loss, and the newsvendor bound itself, with the constants the proof actually supports.

Difficulty

The bound is not a uniform-convergence argument: the class of linear rules on Rp\mathbb R^pRp has complexity growing with ppp, so any bound that holds simultaneously for all rules in the class depends on ppp. The bound must exploit the specific rule produced by the algorithm. The two analytic steps that carry this are the stability of the regularized minimizer, which needs a strong-convexity comparison between the full and the leave-one-out objective, and a concentration inequality of bounded-differences type for a function of the whole sample whose differences are controlled only through stability.

The leave-one-out comparison is a known pitfall. If the leave-one-out problem is run literally on n−1n-1n−1 points it carries the weight 1/(n−1)1/(n-1)1/(n−1), and the comparison argument then gives twice the constant of Theorem 4. The stated constant holds for the leave-one-out objective that keeps the weight 1/n1/n1/n, which is the form of Theorem 5.

Formalization scope

Representation. Features are EuclideanSpace ℝ (Fin p), rules are vectors acting by the inner product, and data points live in EuclideanSpace ℝ (Fin p) × ℝ. The cost is the published newsboy loss with overage cost hhh and underage cost bbb; the empirical and true risks are the published empirical and generalization errors, and the (NV-reg) objective and its 1/n1/n1/n-weighted leave-one-out version are the published regularized objectives of Bousquet and Elisseeff. The regularizer is λ∥q∥22\lambda\|q\|_2^2λ∥q∥22​ (the display prints both λ∥q∥22\lambda\|q\|_2^2λ∥q∥22​ and λ∥q∥2\lambda\|q\|_2λ∥q∥2​; the text calls the problem a quadratic program). "With probability at least 1−δ1-\delta1−δ" is stated as: the event on which the gap exceeds the bound has measure at most δ\deltaδ under the product measure μn\mu^nμn.

Standing assumptions and pinned hypotheses.

  1. b,h,λ>0b,h,\lambda>0b,h,λ>0, Xmax⁡,Dˉ≥0X_{\max},\bar D\ge0Xmax​,Dˉ≥0, n≥1n\ge1n≥1, δ∈(0,1)\delta\in(0,1)δ∈(0,1).
  2. μ\muμ is a probability measure giving full mass to X×[0,Dˉ]\mathcal X\times[0,\bar D]X×[0,Dˉ], with ∥x∥22≤Xmax⁡2\|x\|_2^2\le X_{\max}^2∥x∥22​≤Xmax2​ on X\mathcal XX (§3, p. 9; the page writes the ball as ∥x∥22≤Xmax⁡\|x\|_2^2\le X_{\max}∥x∥22​≤Xmax​, while Theorem 2's "Xmax⁡2X_{\max}^2Xmax2​ as the largest possible value of ∥x∥22\|x\|_2^2∥x∥22​" and Theorem 5's note fix the reading).
  3. The algorithm is any map Sn↦q^(Sn)S_n\mapsto\hat q(S_n)Sn​↦q^​(Sn​) whose value minimizes the (NV-reg) objective, measurable in SnS_nSn​ (Appendix B: "all functions are measurable").
  4. The range assumption of Appendix B: for samples from X×D\mathcal X\times\mathcal DX×D, q^⊤x∈[0,Dˉ]\hat q^\top x\in[0,\bar D]q^​⊤x∈[0,Dˉ] for x∈Xx\in\mathcal Xx∈X.
  5. The last constant is (b∨h)Dˉ(b\vee h)\bar D(b∨h)Dˉ, the loss bound MMM from Lemma 5 that the proof feeds into Theorem 6; display (6) prints Dˉ\bar DDˉ there.
  6. The convention that all sets are countable, and the intercept convention x1=1x^1=1x1=1, are not imposed.

Trivializing formalization ruled out. The range assumption is quantified only over feature vectors in X\mathcal XX and over samples drawn from X×D\mathcal X\times\mathcal DX×D; stated over the whole ball ∥x∥2≤Xmax⁡\|x\|_2\le X_{\max}∥x∥2​≤Xmax​, it would force q^=0\hat q=0q^​=0 (both q^⊤x\hat q^\top xq^​⊤x and q^⊤(−x)\hat q^\top(-x)q^​⊤(−x) would lie in [0,Dˉ][0,\bar D][0,Dˉ]) and the goal would be nearly empty.

What is needed and reusable. McDiarmid's two-sided bounded-differences inequality under product measures; integrability of bounded measurable losses; existence and properties of minimizers of strongly convex objectives on Rp\mathbb R^pRp; the comparison argument behind Theorem 5. Theorem 6 is stated for an arbitrary data space and hypothesis space and is reusable for any uniformly stable algorithm. Contributions to the Theorem 5 reference and to McDiarmid's inequality benefit other missions as well.

Selected references

  • C. Rudin and G.-Y. Vahn, The Big Data Newsvendor: Practical Insights from Machine Learning, MIT Sloan School Working Paper 5036-13, version of February 6, 2014 (MIT DSpace). The version formalized here.
  • G.-Y. Ban and C. Rudin, The Big Data Newsvendor: Practical Insights from Machine Learning, Operations Research 67(1):90–108, 2019. https://doi.org/10.1287/opre.2018.1757
  • O. Bousquet and A. Elisseeff, Stability and Generalization, Journal of Machine Learning Research 2:499–526, 2002. https://jmlr.org/papers/v2/bousquet02a.html
  • C. McDiarmid, On the method of bounded differences, Surveys in Combinatorics, London Math. Soc. Lecture Note Series 141, 148–188, 1989. https://doi.org/10.1017/CBO9781107359949.008
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OptimizationProbabilityStatistics·Captain: mikedeng1

Variance-based Regularization with Convex Objectives III: Localized-Rademacher Risk Bounds for the Robust MinimizerResearch Paper

Why variance-regularized risk bounds

In statistical learning, one picks a function fff from a class F\mathcal FF to make the population risk E[f]\mathbb E[f]E[f] small, with access only to an i.i.d. sample x1,…,xnx_1,\dots,x_nx1​,…,xn​ from an unknown distribution PPP. Empirical risk minimization replaces E[f]\mathbb E[f]E[f] by the empirical mean EP^n[f]\mathbb E_{\widehat P_n}[f]EPn​​[f], and its classical guarantees decay like 1/n1/\sqrt n1/n​ regardless of how concentrated fff is. Bernstein-type inequalities show that the deviation of EP^n[f]\mathbb E_{\widehat P_n}[f]EPn​​[f] from E[f]\mathbb E[f]E[f] scales with the standard deviation of fff, so a procedure that minimizes "empirical risk plus a standard-deviation penalty" can, in principle, achieve faster rates when the variance at the optimum is small (Maurer and Pontil, 2009). The penalized objective is non-convex even when every fff is convex in its parameters, which makes it hard to optimize.

J. C. Duchi and H. Namkoong (arXiv:1610.02581v3, 2017) replace the penalty by a distributionally robust objective: the worst-case risk over all reweightings of the sample within a χ2\chi^2χ2-divergence ball. This objective is convex whenever the losses are, and (Theorem 1 of the paper) it equals the empirical mean plus a standard-deviation penalty up to an error of order 1/n1/n1/n. This mission formalizes the paper's guarantee for the minimizer of that robust objective in terms of localized Rademacher complexities (Section 3.2, Theorem 4), the sharpest of the paper's three generalization analyses. It is the third of four missions on the paper.

Setting

Let PPP be a probability measure on a measurable space X\mathcal XX and x1,…,xnx_1,\dots,x_nx1​,…,xn​, n≥1n\ge1n≥1, an i.i.d. sample from PPP with empirical distribution P^n\widehat P_nPn​. Let M≥1M\ge1M≥1 and let F\mathcal FF be a collection of measurable functions f:X→[0,M]f:\mathcal X\to[0,M]f:X→[0,M] (losses).

  • The χ2\chi^2χ2 ball of radius ρ≥0\rho\ge0ρ≥0 is the set Pn\mathcal P_nPn​ of weight vectors p∈Rnp\in\mathbb R^np∈Rn with pi≥0p_i\ge0pi​≥0, ∑ipi=1\sum_ip_i=1∑i​pi​=1 and 12∑i(npi−1)2≤ρ\frac12\sum_i(np_i-1)^2\le\rho21​∑i​(npi​−1)2≤ρ; equivalently, the distributions PPP on the sample with Dϕ(P∥P^n)≤ρ/nD_\phi(P\|\widehat P_n)\le\rho/nDϕ​(P∥Pn​)≤ρ/n for ϕ(t)=12(t−1)2\phi(t)=\frac12(t-1)^2ϕ(t)=21​(t−1)2.
  • The robust risk of fff is sup⁡P: Dϕ(P∥P^n)≤ρ/nEP[f]=sup⁡p∈Pn∑ipif(xi)\sup_{P:\,D_\phi(P\|\widehat P_n)\le\rho/n}\mathbb E_P[f]=\sup_{p\in\mathcal P_n}\sum_ip_if(x_i)supP:Dϕ​(P∥Pn​)≤ρ/n​EP​[f]=supp∈Pn​​∑i​pi​f(xi​), and a robust minimizer f^\widehat ff​ minimizes it over F\mathcal FF.
  • The empirical Rademacher complexity is Rn(F)=Eε[sup⁡f∈F1n∑iεif(xi)]\mathfrak R_n(\mathcal F)=\mathbb E_\varepsilon\big[\sup_{f\in\mathcal F}\frac1n\sum_i\varepsilon_if(x_i)\big]Rn​(F)=Eε​[supf∈F​n1​∑i​εi​f(xi​)] with i.i.d. uniform signs εi∈{−1,1}\varepsilon_i\in\{-1,1\}εi​∈{−1,1}, and E[Rn(F)]\mathbb E[\mathfrak R_n(\mathcal F)]E[Rn​(F)] averages it over the sample.
  • A function ψ:R+→R+\psi:\mathbb R_+\to\mathbb R_+ψ:R+​→R+​ is sub-root if it is nonnegative, nondecreasing, and r↦ψ(r)/rr\mapsto\psi(r)/\sqrt rr↦ψ(r)/r​ is nonincreasing on r>0r>0r>0.
  • The localization inequality (20) asks that, for all r≥0r\ge0r≥0,
ψn(r) ≥ E[Rn({cf:f∈F, c∈[0,1], E[c2f2]≤r})],\psi_n(r)\ \ge\ \mathbb E\big[\mathfrak R_n(\{cf : f\in\mathcal F,\ c\in[0,1],\ \mathbb E[c^2f^2]\le r\})\big],ψn​(r) ≥ E[Rn​({cf:f∈F, c∈[0,1], E[c2f2]≤r})],

with ψn\psi_nψn​ sub-root, and rn⋆>0r_n^\star>0rn⋆​>0 is a point with rn⋆≥ψn(rn⋆)r_n^\star\ge\psi_n(r_n^\star)rn⋆​≥ψn​(rn⋆​).

Formalization targets

Goal: Theorem 4, inequality (23), as its proof establishes it

Let 0<t<n0<t<n0<t<n and let ρ\rhoρ satisfy (21): ρn≥8(45Mn(t+log⁡⌈log⁡nt⌉)+18rn⋆)\frac\rho n\ge8\big(\frac{45M}n\big(t+\log\lceil\log\frac nt\rceil\big)+18r_n^\star\big)nρ​≥8(n45M​(t+log⌈logtn​⌉)+18rn⋆​). With probability at least 1−4e−t1-4e^{-t}1−4e−t, every robust minimizer f^\widehat ff​ satisfies

E[f^] ≤ (1+22ρn)inf⁡f∈F(E[f]+182ρ45nVar(f))+(14+62ρn)M(3ρ+t)n.\mathbb E[\widehat f]\ \le\ \Big(1+2\sqrt{\tfrac{2\rho}n}\Big)\inf_{f\in\mathcal F}\Big(\mathbb E[f]+\sqrt{\tfrac{182\rho}{45n}\mathrm{Var}(f)}\Big)+\Big(14+6\sqrt{\tfrac{2\rho}n}\Big)\frac{M(3\rho+t)}n .E[f​] ≤ (1+2n2ρ​​)f∈Finf​(E[f]+45n182ρ​Var(f)​)+(14+6n2ρ​​)nM(3ρ+t)​.

Milestones

In attack order: Bousquet's form of Talagrand's inequality (Lemma B.2); the elementary root bound (Lemma D.4); the contraction principle (Lemma D.5, a published theorem); the uniform Bernstein inequality with Rademacher complexity (Lemma D.1); its localized version in terms of rn⋆r_n^\starrn⋆​ (Lemma D.2); localized second-moment bounds (Lemma D.3); the deterministic expansion (10) of Theorem 1,

(2ρnsn2−2Mρn)+≤sup⁡PEP[Z]−EP^n[Z]≤2ρnsn2;\Big(\sqrt{\tfrac{2\rho}n s_n^2}-\tfrac{2M\rho}n\Big)_+\le\sup_{P}\mathbb E_P[Z]-\mathbb E_{\widehat P_n}[Z]\le\sqrt{\tfrac{2\rho}ns_n^2};(n2ρ​sn2​​−n2Mρ​)+​≤Psup​EP​[Z]−EPn​​[Z]≤n2ρ​sn2​​;

and the uniform bound (22): with probability at least 1−2e−t1-2e^{-t}1−2e−t, for all f∈Ff\in\mathcal Ff∈F,

E[f]≤(1+22ρn)sup⁡P: Dϕ(P∥P^n)≤ρ/nEP[f]+(13+42ρn)Mρn.\mathbb E[f]\le\Big(1+2\sqrt{\tfrac{2\rho}n}\Big)\sup_{P:\,D_\phi(P\|\widehat P_n)\le\rho/n}\mathbb E_P[f]+\Big(13+4\sqrt{\tfrac{2\rho}n}\Big)\frac{M\rho}n .E[f]≤(1+2n2ρ​​)P:Dϕ​(P∥Pn​)≤ρ/nsup​EP​[f]+(13+4n2ρ​​)nMρ​.

Significance

The bound (23) says that the robust minimizer competes with the best trade-off between risk and standard deviation in the class, and that the complexity of the class enters only through the fixed point rn⋆r_n^\starrn⋆​ of a localized complexity bound. For bounded VC classes rn⋆r_n^\starrn⋆​ is of order dlog⁡(n/d)n\frac{d\log(n/d)}nndlog(n/d)​ (Bartlett, Bousquet and Mendelson, 2005, Corollary 3.7), so when the optimal function has small variance the excess risk is of order ρ/n\rho/nρ/n, faster than the 1/n1/\sqrt n1/n​ of uniform covering arguments; and localized complexities apply to classes, such as balls of reproducing kernel Hilbert spaces, whose covering numbers are too large for the covering-number analysis of the paper's Theorem 3 (mission II of this series).

The paper's result is proved, not open. No part of it, and none of the localized-complexity machinery of Bartlett, Bousquet and Mendelson, is formalized in Lean or Mathlib to our knowledge. The mission produces a checked version of the theorem with every constant explicit and, along the way, the localization lemmas D.1–D.3, which are reusable for any localized-complexity analysis. Reading the proof also exposed three arithmetic slips in the printed statements; the mission states what the proof establishes (see Formalization scope).

Difficulty

The obvious route applies a uniform concentration inequality to F\mathcal FF and then a Bernstein bound to each fff. Talagrand's inequality applied to the whole class gives a deviation governed by the largest variance in the class and by the global complexity E[Rn(F)]\mathbb E[\mathfrak R_n(\mathcal F)]E[Rn​(F)], which yields only 1/n1/\sqrt n1/n​ rates. Obtaining a deviation that scales with each function's own second moment requires peeling the class into shells of comparable second moment and a fixed-point argument on the sub-root bound, with a union bound whose cost appears as log⁡⌈log⁡nt⌉\log\lceil\log\frac nt\rceillog⌈logtn​⌉. The two directions of the localized inequalities (population to sample, and sample to population for second moments) must then be combined with the deterministic expansion (10) while keeping the constants explicit. A further subtlety is the self-normalized rescaling f↦r/(E[f2]∨r) ff\mapsto\sqrt{r/(\mathbb E[f^2]\vee r)}\,ff↦r/(E[f2]∨r)​f, which differs from the variance normalization of Bartlett et al. and is what makes the bound compatible with the robust objective.

Formalization scope

Lean conventions. The sample is the coordinate map of the product measure PnP^nPn on Fin n → X. Distributions on the sample are weight vectors in the χ2\chi^2χ2 ball; the robust risk is the real supremum over that ball (attained, since the ball is nonempty and compact for n≥1n\ge1n≥1, ρ≥0\rho\ge0ρ≥0). Population means and variances are ∫ x, f x ∂P and ProbabilityTheory.variance f P for measurable bounded fff; empirical means and variances are normalized by 1/n1/n1/n. The empirical Rademacher complexity is the published UnderstandingML_Rademacher definition evaluated on {(f(x1),…,f(xn))}\{(f(x_1),\dots,f(x_n))\}{(f(x1​),…,f(xn​))}. Its expectation is a Bochner integral, and every hypothesis that bounds it also asserts that the integrand is integrable: otherwise the integral is 000, (20) would hold for free, and the theorem would be false. Probability bounds are stated for the failure event under PnP^nPn (an outer measure when the event is not measurable). The goal speaks about every minimizer of the robust risk, so it is not vacuous when the set of minimizers is empty. The condition rn⋆>0r_n^\star>0rn⋆​>0 is part of the page's "root" (and the proof divides by rn⋆\sqrt{r_n^\star}rn⋆​​); with rn⋆=0r_n^\star=0rn⋆​=0 allowed, ψ(r)=r\psi(r)=\sqrt rψ(r)=r​ would remove the complexity term from (21). The condition t<nt<nt<n makes log⁡⌈log⁡nt⌉\log\lceil\log\frac nt\rceillog⌈logtn​⌉ defined.

Corrections of printed statements, each recorded in the item's docstring and Formalization Note (the milestone texts stay verbatim):

  • (22) is stated with probability 1−2e−t1-2e^{-t}1−2e−t; the paper prints 1−e−t1-e^{-t}1−e−t, and its proof (p. 41) concludes 1−2e−t1-2e^{-t}1−2e−t.
  • (23) is stated with probability 1−4e−t1-4e^{-t}1−4e−t (printed 1−3e−t1-3e^{-t}1−3e−t; the proof adds two fixed-fff events to the two of (22)) and with 182ρ45n\frac{182\rho}{45n}45n182ρ​ (printed 91ρ45n\frac{91\rho}{45n}45n91ρ​; the proof's step ρ+t≤91ρ/45\sqrt\rho+\sqrt t\le\sqrt{91\rho/45}ρ​+t​≤91ρ/45​ multiplies 2Var(f)/n\sqrt{2\mathrm{Var}(f)/n}2Var(f)/n​).
  • Lemma D.3 is stated with the additive term 72M2(1+η)rn⋆+(4(1+η)+143)M2tn72M^2(1+\eta)r_n^\star+(4(1+\eta)+\frac{14}3)\frac{M^2t}n72M2(1+η)rn⋆​+(4(1+η)+314​)nM2t​ and, in the reversed direction, the coefficient 1+11+η1+\frac1{1+\eta}1+1+η1​, as its proof yields (printed: Mtn(4+73M)\frac{Mt}n(4+\frac73M)nMt​(4+37​M) and 1+η1+η1+\frac\eta{1+\eta}1+1+ηη​), under Theorem 4's standing hypothesis M≥1M\ge1M≥1.
  • Lemma D.5 is linked to the published contraction lemma UnderstandingML.contraction_lemma, which states it at a fixed sample for nonempty bounded classes and allows a different Lipschitz map per coordinate.

Contributions welcome: proofs of the milestones in any order; Lemma B.2 (Bousquet's inequality) is the deepest single ingredient and is reusable well beyond this mission, as are the peeling Lemma D.1 and the sub-root fixed-point Lemma D.2.

Selected references

  • J. C. Duchi and H. Namkoong, Variance-based regularization with convex objectives, arXiv:1610.02581v3, 2017. https://arxiv.org/abs/1610.02581
  • P. L. Bartlett, O. Bousquet and S. Mendelson, Local Rademacher complexities, Annals of Statistics 33(4), 2005. https://doi.org/10.1214/009053605000000282
  • O. Bousquet, A Bennett concentration inequality and its application to suprema of empirical processes, Comptes Rendus Mathématique 334(6), 2002. https://doi.org/10.1016/S1631-073X(02)02292-6
  • A. Maurer and M. Pontil, Empirical Bernstein bounds and sample variance penalization, COLT 2009. https://arxiv.org/abs/0907.3740
  • M. Ledoux and M. Talagrand, Probability in Banach Spaces, Springer, 1991. https://doi.org/10.1007/978-3-642-20212-4
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Bandit AlgorithmsOperations ResearchProbability·Captain: mikedeng1

Analysis of Thompson Sampling for the Multi-armed Bandit Problem 2: Logarithmic Regret for N ArmsResearch Paper

Motivation

Thompson Sampling is the oldest heuristic for the multi-armed bandit problem: proposed by Thompson in 1933, it plays each arm with the posterior probability that the arm is the best one. It is simple to implement, performs well empirically (Chapelle and Li, NIPS 2011), and has been used in production systems such as click-through-rate prediction for search advertising. For a long time, however, no finite-time regret guarantee was known for it: the analyses available before 2012 gave only o(T)o(T)o(T) regret.

Agrawal and Goyal (arXiv:1111.1797, COLT 2012) gave the first logarithmic bounds on the expected regret of Thompson Sampling. This mission formalizes their bound for the general case of NNN arms (their Theorem 2). A companion mission of the same series formalizes their two-armed bound (Theorem 1), whose proof is independent.

Timeline. Lai and Robbins (1985) proved that every consistent algorithm has regret at least of order ∑iΔiD(μi∥μ1)ln⁡T\sum_i \frac{\Delta_i}{D(\mu_i\|\mu_1)}\ln T∑i​D(μi​∥μ1​)Δi​​lnT. Auer, Cesa-Bianchi and Fischer (2002) showed that UCB1 achieves O(∑iln⁡T/Δi)O(\sum_i \ln T/\Delta_i)O(∑i​lnT/Δi​) in finite time. Agrawal and Goyal (2012) proved O((∑a1/Δa2)2ln⁡T)O((\sum_a 1/\Delta_a^2)^2\ln T)O((∑a​1/Δa2​)2lnT) for Thompson Sampling with NNN arms; Kaufmann, Korda and Munos (2012) and Agrawal and Goyal (2013) later proved the asymptotically optimal constant for Bernoulli rewards.

Setting

A stochastic NNN-armed bandit has arms 1,…,N1,\dots,N1,…,N. Arm iii, when played, yields a random reward drawn from a fixed distribution νi\nu_iνi​ supported in [0,1][0,1][0,1], with mean μi\mu_iμi​; rewards of an arm are i.i.d. and independent of the other arms. Arm 111 is assumed to be the unique optimal arm, μ1>μi\mu_1>\mu_iμ1​>μi​ for i≠1i\ne1i=1, and Δi=μ1−μi>0\Delta_i=\mu_1-\mu_i>0Δi​=μ1​−μi​>0 is the gap of arm iii.

Thompson Sampling for general stochastic bandits (Algorithm 2 of the paper) keeps, for each arm iii, a count SiS_iSi​ of successes and FiF_iFi​ of failures, both starting at 000. In each round ttt it draws θi(t)∼Beta(Si+1,Fi+1)\theta_i(t)\sim\mathrm{Beta}(S_i+1,F_i+1)θi​(t)∼Beta(Si​+1,Fi​+1) independently for every arm, plays i(t)=arg⁡max⁡iθi(t)i(t)=\arg\max_i\theta_i(t)i(t)=argmaxi​θi​(t), observes a reward r~t∼νi(t)\tilde r_t\sim\nu_{i(t)}r~t​∼νi(t)​, performs a Bernoulli trial with success probability r~t\tilde r_tr~t​, and increments Si(t)S_{i(t)}Si(t)​ on success and Fi(t)F_{i(t)}Fi(t)​ on failure.

The expected regret in time TTT is

E[R(T)]=E[∑t=1T(μ∗−μi(t))],μ∗=max⁡iμi,\mathbb E[\mathcal R(T)]=\mathbb E\Big[\sum_{t=1}^T(\mu^*-\mu_{i(t)})\Big],\qquad \mu^*=\max_i\mu_i,E[R(T)]=E[t=1∑T​(μ∗−μi(t)​)],μ∗=imax​μi​,

the expectation being over the rewards, the Bernoulli trials and the posterior samples.

The proof works with the reward stacks Zi,mZ_{i,m}Zi,m​: the outcome of the mmm-th Bernoulli trial of arm iii, all independent. Then s(j)=∑m≤jZ1,ms(j)=\sum_{m\le j}Z_{1,m}s(j)=∑m≤j​Z1,m​, the number of successes in the first jjj plays of arm 111, is a Binomial(j,μ1)\mathrm{Binomial}(j,\mu_1)Binomial(j,μ1​) random variable. The other objects of the proof are the threshold Li=24ln⁡T/Δi2L_i=24\ln T/\Delta_i^2Li​=24lnT/Δi2​, the saturated set C(t)C(t)C(t) of suboptimal arms with at least LiL_iLi​ plays before round ttt, the intervals IjI_jIj​ between the jjj-th and (j+1)(j+1)(j+1)-th plays of arm 111, and the counts γj\gamma_jγj​ and Vjℓ,aV_j^{\ell,a}Vjℓ,a​ defined in §4.

Formalization targets

Goal: Theorem 2

There is an absolute constant C>0C>0C>0 such that for every N≥2N\ge2N≥2, every instance as above and every horizon T≥2T\ge2T≥2,

E[R(T)]≤C(∑a=2N1Δa2)2ln⁡T.\mathbb E[\mathcal R(T)]\le C\Big(\sum_{a=2}^N\frac{1}{\Delta_a^2}\Big)^2\ln T .E[R(T)]≤C(a=2∑N​Δa2​1​)2lnT.

CCC does not depend on NNN, on the reward distributions or on TTT.

Milestones

  1. Lemma 4: with E(t)E(t)E(t) the event that every saturated arm's sample lies within Δi/2\Delta_i/2Δi​/2 of its mean, Pr⁡(E(t))≥1−4(N−1)/T2\Pr(E(t))\ge1-4(N-1)/T^2Pr(E(t))≥1−4(N−1)/T2, also conditionally on s(j)=ss(j)=ss(j)=s.
  2. Lemma 5 (Eq. (7)): the expected regret from saturated arms inside IjI_jIj​ is at most E[E[γj+1∣s(j)]∑aΔaE[min⁡{X(j,s(j),μa+Δa/2),T}∣s(j)]]\mathbb E\big[\mathbb E[\gamma_j+1\mid s(j)]\sum_a\Delta_a\mathbb E[\min\{X(j,s(j),\mu_a+\Delta_a/2),T\}\mid s(j)]\big]E[E[γj​+1∣s(j)]∑a​Δa​E[min{X(j,s(j),μa​+Δa​/2),T}∣s(j)]].
  3. Lemma 1: E[X(j,s,y)]=1/Fj+1,yB(s)−1\mathbb E[X(j,s,y)]=1/F^B_{j+1,y}(s)-1E[X(j,s,y)]=1/Fj+1,yB​(s)−1, where X(j,s,y)X(j,s,y)X(j,s,y) counts the trials before an independent Beta(s+1,j−s+1)\mathrm{Beta}(s+1,j-s+1)Beta(s+1,j−s+1) sample exceeds yyy.
  4. Lemma 3: a three-case bound on E[E[min⁡{X(j,s(j),y),T}∣s(j)]]\mathbb E[\mathbb E[\min\{X(j,s(j),y),T\}\mid s(j)]]E[E[min{X(j,s(j),y),T}∣s(j)]] in terms of the Bernoulli KL divergence DDD between yyy and μ1\mu_1μ1​.

Significance

The result. Theorem 2 shows that Thompson Sampling, a randomized Bayesian heuristic, achieves regret logarithmic in the horizon for any number of arms with bounded rewards, matching the order in TTT of the Lai–Robbins lower bound. Its dependence on the gaps, (∑aΔa−2)2(\sum_a\Delta_a^{-2})^2(∑a​Δa−2​)2, is worse than UCB1's; the paper's own Remark 1 and later work improve it. The proof introduced the device of bounding the waiting time between plays of the optimal arm through geometric variables with Beta-cdf parameters (Lemmas 1 and 3), which reappears in later analyses of Thompson Sampling.

Formalizing it. The theorem is proved on paper; it has not been machine-checked. Bandit theory in Lean (bandit environments, regret, UCB-type analyses) is still young, and no Beta–Bernoulli Thompson Sampling result is formalized. The mission produces a Lean model of Algorithm 2 for general [0,1][0,1][0,1] rewards with the paper's stack coupling, the §4 bookkeeping of saturated arms and intervals, and the paper's lemmas as separate targets.

Difficulty

Two difficulties are specific to the NNN-armed analysis. First, the arm that competes with arm 111 changes over time: the set of saturated arms grows, and which saturated arm is "best" depends on the history, so the waiting time between plays of arm 111 cannot be compared with a single geometric variable as in the two-armed case. Second, the number γj\gamma_jγj​ of rounds at which arm 111's sample is large but arm 111 is not played is not independent of the counts Vjℓ,aV_j^{\ell,a}Vjℓ,a​: both depend on the same posterior samples, and Lemma 5 needs a careful conditioning on the history to separate them. The obvious union bound over arms, treating each suboptimal arm as in the two-armed proof, fails because it ignores the interruptions by unsaturated arms, whose number is the source of the squared sum in the bound.

Formalization scope

  • Probability space. Algorithm 2 is realized on a product of three independent i.i.d. tables: Beta draws indexed by (arm, round, successes, failures), rewards indexed by (arm, round) and uniform variables indexed by (arm, round); the Bernoulli trial of a round succeeds when the played arm's uniform variable is below its reward. The law of the run is that of Algorithm 2, which runs for every round t=1,2,…t=1,2,\dotst=1,2,…. s(j)s(j)s(j) is the number of successful trials among the first jjj plays of arm 111 in this infinite run (possibly after the horizon TTT), so it is a Binomial(j,μ1)\mathrm{Binomial}(j,\mu_1)Binomial(j,μ1​) random variable for every jjj, as the paper's independent Z1,mZ_{1,m}Z1,m​ make it. Ties in the arg max (probability 000) go to the smallest index.
  • Indexing. Arms are Fin N, and Lean arm 0 is the paper's arm 111. Rounds are 0,…,T−10,\dots,T-10,…,T−1; Lean round ttt is the paper's round t+1t+1t+1. Sums over a=2,…,Na=2,\dots,Na=2,…,N are sums over a≠0a\ne0a=0.
  • Expectations are lower Lebesgue integrals in [0,∞][0,\infty][0,∞], which has no junk value for non-integrable functions. Conditional expectations given s(j)s(j)s(j) are written as finite sums over the values of s(j)s(j)s(j).
  • The O(⋅)O(\cdot)O(⋅). The paper writes O(⋅)O(\cdot)O(⋅) in the sense of its footnote 1 (f(n)≤c g(n)f(n)\le c\,g(n)f(n)≤cg(n) for n≥n0n\ge n_0n≥n0​). The goal states it with one universal constant CCC, quantified before NNN, the instance and TTT, for every T≥2T\ge2T≥2. The explicit constants printed in App. D are not formalized: expanding the paper's Eq. (21) gives terms 288(N−1)(ln⁡T)∑aΔa−2288(N-1)(\ln T)\sum_a\Delta_a^{-2}288(N−1)(lnT)∑a​Δa−2​ and 48(N−1)248(N-1)^248(N−1)2 where the paper prints 288(ln⁡T)∑iΔi−2288(\ln T)\sum_i\Delta_i^{-2}288(lnT)∑i​Δi−2​, and Eq. (22) drops a factor ln⁡T\ln TlnT in its 192/Δa2192/\Delta_a^2192/Δa2​ term. The O(⋅)O(\cdot)O(⋅) claim does not depend on these slips; a statement pinned to the printed numerals might be false.
  • Ruled out. A constant depending on NNN, on the means or on TTT; a fixed number of arms; Bernoulli rewards only; or any algorithm other than Algorithm 2 would each make the goal a different and weaker theorem. The statement quantifies over all N≥2N\ge2N≥2 and all reward distributions on [0,1][0,1][0,1].
  • Not included. Eq. (8), the bound ∑jE[γj∣s(j)]≤∑uLu+4(N−1)\sum_{j}\mathbb E[\gamma_j\mid s(j)]\le\sum_uL_u+4(N-1)∑j​E[γj​∣s(j)]≤∑u​Lu​+4(N−1) "for all instantiations", is not a milestone: each term is conditioned on a different s(j)s(j)s(j), and the pointwise reading does not follow from the argument given. Remark 1 (an alternate bound) and App. A (several optimal arms) are not part of this mission.
  • Contributions welcome: Beta–Binomial identities, geometric waiting times, Hoeffding bounds for binomial cdfs, and the stopping-time arguments behind Lemma 5. Lemma 1 and Lemma 3 are shared with the two-armed mission of this series.

Selected references

  • S. Agrawal and N. Goyal, Analysis of Thompson Sampling for the Multi-armed Bandit Problem, COLT 2012; arXiv:1111.1797v3, 2012. https://arxiv.org/abs/1111.1797
  • W. R. Thompson, On the likelihood that one unknown probability exceeds another in view of the evidence of two samples, Biometrika 25, 1933. https://doi.org/10.1093/biomet/25.3-4.285
  • T. L. Lai and H. Robbins, Asymptotically efficient adaptive allocation rules, Advances in Applied Mathematics 6, 1985. https://doi.org/10.1016/0196-8858(85)90002-8
  • P. Auer, N. Cesa-Bianchi and P. Fischer, Finite-time analysis of the multiarmed bandit problem, Machine Learning 47, 2002. https://doi.org/10.1023/A:1013689704352
  • O. Chapelle and L. Li, An empirical evaluation of Thompson Sampling, NIPS 2011. https://papers.nips.cc/paper/4321-an-empirical-evaluation-of-thompson-sampling
  • E. Kaufmann, N. Korda and R. Munos, Thompson Sampling: an asymptotically optimal finite-time analysis, ALT 2012. https://arxiv.org/abs/1205.4217
  • S. Agrawal and N. Goyal, Further optimal regret bounds for Thompson Sampling, AISTATS 2013. https://arxiv.org/abs/1209.3353
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Linear algebraOptimizationReinforcement Learning·Captain: mikedeng1

Reinforcement Learning: An Introduction X: Off-policy Divergence and the Gradient of the Projected Bellman ErrorTextbook

Motivation

Off-policy learning estimates the value function of a target policy π\piπ from data generated by a different behavior policy bbb. It is how an agent learns about a greedy policy while exploring, and how many policies can be evaluated from one stream of experience. Combined with linear function approximation and bootstrapping (updating an estimate toward other estimates, as temporal-difference methods do), off-policy learning can be unstable: the weights can diverge even when every quantity involved is well defined. Sutton and Barto call this combination the deadly triad (Chapter 11 of Reinforcement Learning: An Introduction, 2nd ed., 2018).

Chapter 11 does two things. It exhibits the instability with small, fully computable examples, and it identifies an objective that can be minimized stably from off-policy data: the mean square projected Bellman error (PBE), whose gradient the Gradient-TD methods (GTD2, TDC) follow in expectation. This mission formalizes the chapter's exact, finite-dimensional claims.

A short history, following the book's bibliographical remarks (pp. 285–286): Baird (1995) gave the seven-state counterexample for off-policy semi-gradient TD; Tsitsiklis and Van Roy (1996) gave the earliest w-to-2w example and the counterexample of Example 11.1, showing that even a least-squares fit at every step can diverge; Gradient-TD methods, which follow the gradient of the PBE, were introduced by Sutton, Szepesvári and Maei and by Sutton et al. (2009); Sutton, Mahmood and White (2016) introduced Emphatic-TD. The learnability discussion of §11.6 is the book's own.

Setting

A finite Markov decision process has states S\mathcal SS, actions A\mathcal AA, a finite reward set R\mathcal RR and dynamics p(s′,r∣s,a)p(s',r\mid s,a)p(s′,r∣s,a). A policy π(a∣s)\pi(a\mid s)π(a∣s) induces the transition matrix Pπ(s,s′)=∑aπ(a∣s) p(s′∣s,a)P_\pi(s,s') = \sum_a \pi(a\mid s)\,p(s'\mid s,a)Pπ​(s,s′)=∑a​π(a∣s)p(s′∣s,a) and expected reward rπ(s)r_\pi(s)rπ​(s). For 0≤γ<10\le\gamma<10≤γ<1 the true value function is the expected discounted return vπ(s)=∑k≥0γk(Pπkrπ)(s)v_\pi(s) = \sum_{k\ge0}\gamma^k (P_\pi^k r_\pi)(s)vπ​(s)=∑k≥0​γk(Pπk​rπ​)(s).

A state weighting μ\muμ is a probability distribution on S\mathcal SS, with D=diag⁡(μ)\mathbf D = \operatorname{diag}(\mu)D=diag(μ) and norm ∥v∥μ2=∑sμ(s)v(s)2\|v\|_\mu^2 = \sum_s \mu(s)v(s)^2∥v∥μ2​=∑s​μ(s)v(s)2. The ∣S∣×d|\mathcal S|\times d∣S∣×d matrix X\mathbf XX has the feature vectors x(s)⊤\mathbf x(s)^\topx(s)⊤ as rows, and a weight vector w∈Rd\mathbf w\in\mathbb R^dw∈Rd gives the linear value function vw=Xwv_{\mathbf w} = \mathbf X\mathbf wvw​=Xw. The Bellman operator is

(Bπv)(s)=∑aπ(a∣s)∑s′,rp(s′,r∣s,a) [r+γv(s′)].(B_\pi v)(s) = \sum_a \pi(a\mid s)\sum_{s',r} p(s',r\mid s,a)\,[r+\gamma v(s')].(Bπ​v)(s)=a∑​π(a∣s)s′,r∑​p(s′,r∣s,a)[r+γv(s′)].

The Bellman error vector is δˉw=Bπvw−vw\bar\delta_{\mathbf w} = B_\pi v_{\mathbf w} - v_{\mathbf w}δˉw​=Bπ​vw​−vw​, the projection is Π=X(X⊤DX)−1X⊤D\Pi = \mathbf X(\mathbf X^\top\mathbf D\mathbf X)^{-1}\mathbf X^\top\mathbf DΠ=X(X⊤DX)−1X⊤D, and the two objectives are

BE(w)=∥δˉw∥μ2,PBE(w)=∥Πδˉw∥μ2.\mathrm{BE}(\mathbf w) = \|\bar\delta_{\mathbf w}\|_\mu^2,\qquad \mathrm{PBE}(\mathbf w) = \|\Pi\bar\delta_{\mathbf w}\|_\mu^2 .BE(w)=∥δˉw​∥μ2​,PBE(w)=∥Πδˉw​∥μ2​.

Off-policy samples are reweighted by the importance-sampling ratio ρt=π(At∣St)/b(At∣St)\rho_t = \pi(A_t\mid S_t)/b(A_t\mid S_t)ρt​=π(At​∣St​)/b(At​∣St​).

Formalization targets

Goal: the gradient of the PBE

If X⊤DX\mathbf X^\top\mathbf D\mathbf XX⊤DX is invertible, then for every w\mathbf ww

PBE(w)=(X⊤Dδˉw)⊤(X⊤DX)−1(X⊤Dδˉw),\mathrm{PBE}(\mathbf w) = (\mathbf X^\top\mathbf D\bar\delta_{\mathbf w})^\top(\mathbf X^\top\mathbf D\mathbf X)^{-1}(\mathbf X^\top\mathbf D\bar\delta_{\mathbf w}),PBE(w)=(X⊤Dδˉw​)⊤(X⊤DX)−1(X⊤Dδˉw​), ∇PBE(w)=2 (γPπX−X)⊤DX (X⊤DX)−1 X⊤Dδˉw.\nabla\mathrm{PBE}(\mathbf w) = 2\,(\gamma P_\pi\mathbf X-\mathbf X)^\top\mathbf D\mathbf X\,(\mathbf X^\top\mathbf D\mathbf X)^{-1}\,\mathbf X^\top\mathbf D\bar\delta_{\mathbf w}.∇PBE(w)=2(γPπ​X−X)⊤DX(X⊤DX)−1X⊤Dδˉw​.

With μ\muμ the state distribution under bbb this is (11.27), ∇PBE(w)=2 E[ρt(γxt+1−xt)xt⊤] E[xtxt⊤]−1 E[ρtδtxt]\nabla\mathrm{PBE}(\mathbf w) = 2\,\mathbb E[\rho_t(\gamma\mathbf x_{t+1}-\mathbf x_t)\mathbf x_t^\top]\,\mathbb E[\mathbf x_t\mathbf x_t^\top]^{-1}\,\mathbb E[\rho_t\delta_t\mathbf x_t]∇PBE(w)=2E[ρt​(γxt+1​−xt​)xt⊤​]E[xt​xt⊤​]−1E[ρt​δt​xt​].

Milestones, in attack order

  1. The w-to-2w example (p. 260): repeated off-policy semi-gradient TD(0) on one transition multiplies www by 1+α(2γ−1)1+\alpha(2\gamma-1)1+α(2γ−1), so wt→±∞w_t\to\pm\inftywt​→±∞ for every α>0\alpha>0α>0 once γ>12\gamma>\tfrac12γ>21​.
  2. Example 11.1, (11.10) (p. 263): the least-squares iteration wk+1=6−4ε5γwkw_{k+1} = \tfrac{6-4\varepsilon}{5}\gamma w_kwk+1​=56−4ε​γwk​ diverges when γ>5/(6−4ε)\gamma>5/(6-4\varepsilon)γ>5/(6−4ε) and w0≠0w_0\ne0w0​=0.
  3. (11.12)–(11.13): Πv\Pi vΠv is the unique μ\muμ-closest representable function, and Π⊤DΠ=DX(X⊤DX)−1X⊤D\Pi^\top\mathbf D\Pi = \mathbf D\mathbf X(\mathbf X^\top\mathbf D\mathbf X)^{-1}\mathbf X^\top\mathbf DΠ⊤DΠ=DX(X⊤DX)−1X⊤D.
  4. (11.21): vπv_\pivπ​ is the unique fixed point of BπB_\piBπ​ (γ<1\gamma<1γ<1).
  5. (11.24), Exercise 11.4: RE(w)=VE(w)+E[(Gt−vπ(St))2]\mathrm{RE}(\mathbf w) = \mathrm{VE}(\mathbf w) + \mathbb E[(G_t - v_\pi(S_t))^2]RE(w)=VE(w)+E[(Gt​−vπ​(St​))2] in the on-policy case.
  6. Example 11.4 (p. 276): two Markov reward processes that generate the same observable data distribution (every finite prefix of the stream of feature vectors and rewards has the same probability, each process started from its stationary distribution) have BE(0)=0\mathrm{BE}(\mathbf 0) = 0BE(0)=0 and BE(0)=23\mathrm{BE}(\mathbf 0) = \tfrac23BE(0)=32​.
  7. (11.25)–(11.26): the PBE in matrix terms.
  8. The three factors (p. 278): X⊤Dδˉw=E[ρtδtxt]\mathbf X^\top\mathbf D\bar\delta_{\mathbf w} = \mathbb E[\rho_t\delta_t\mathbf x_t]X⊤Dδˉw​=E[ρt​δt​xt​], (γPπX−X)⊤DX=E[ρt(γxt+1−xt)xt⊤](\gamma P_\pi\mathbf X-\mathbf X)^\top\mathbf D\mathbf X = \mathbb E[\rho_t(\gamma\mathbf x_{t+1}-\mathbf x_t)\mathbf x_t^\top](γPπ​X−X)⊤DX=E[ρt​(γxt+1​−xt​)xt⊤​], X⊤DX=E[xtxt⊤]\mathbf X^\top\mathbf D\mathbf X = \mathbb E[\mathbf x_t\mathbf x_t^\top]X⊤DX=E[xt​xt⊤​], under coverage.

Significance

The results. The divergence examples show that no step-size choice rescues semi-gradient TD off-policy, and that exact least-squares fitting does not either. The learnability results separate objectives that can be estimated from observed features and rewards (RE, PBE) from one that cannot (BE), and (11.24) shows that the unobservable VE has the same minimizer as the observable RE. The gradient formula (11.27) is the expected update of the Gradient-TD family; its factorization into three expectations is what makes an O(d)O(d)O(d) stochastic-gradient method possible.

Formalizing them. All of these are proved or computed in the book, in informal notation that leaves hypotheses implicit (invertibility, coverage, the distribution μ\muμ, the domain of γ\gammaγ). A machine-checked version fixes them, and gives a reusable layer of linear value-function geometry (Bellman operator, μ\muμ-norm, projection, BE, PBE, behavior-policy expectations) for later work on Gradient-TD convergence and on the TD fixed point. No formalization of this chapter is known to exist.

Difficulty

The individual steps are finite linear algebra, but they are easy to get wrong. The gradient of a quadratic form h⊤C−1h\mathbf h^\top\mathbf C^{-1}\mathbf hh⊤C−1h in an affine h(w)\mathbf h(\mathbf w)h(w) produces C−1+C−⊤\mathbf C^{-1}+\mathbf C^{-\top}C−1+C−⊤, and collapsing it to 2C−12\mathbf C^{-1}2C−1 uses the symmetry of X⊤DX\mathbf X^\top\mathbf D\mathbf XX⊤DX. The Jacobian of w↦X⊤Dδˉw\mathbf w\mapsto\mathbf X^\top\mathbf D\bar\delta_{\mathbf w}w↦X⊤Dδˉw​ has to be identified through the Bellman operator's affine form rπ+γPπvr_\pi+\gamma P_\pi vrπ​+γPπ​v, which is a theorem about the four-argument dynamics, not a definition. Passing from matrix to expectation form needs coverage, since otherwise b(a∣s)ρ=π(a∣s)b(a\mid s)\rho = \pi(a\mid s)b(a∣s)ρ=π(a∣s) fails. The divergence examples need the conclusion ∣wk∣→∞|w_k|\to\infty∣wk​∣→∞, not merely "the multiplier exceeds one".

Formalization scope

States are a finite type; values are functions S→R\mathcal S\to\mathbb RS→R; weights are Rd\mathbb R^dRd as Fin d → ℝ; matrices are Mathlib Matrix. The dynamics are the book's p(s′,r∣s,a)p(s',r\mid s,a)p(s′,r∣s,a) with a finite reward set, and one action type serves all states. μ\muμ is a probability distribution (nonnegative, summing to one). vπv_\pivπ​ is defined from expected returns, not as a Bellman solution. The gradient is a Fréchet derivative whose linear map is u↦g⊤u\mathbf u\mapsto\mathbf g^\top\mathbf uu↦g⊤u.

Explicit hypotheses the book leaves implicit:

  • X⊤DX\mathbf X^\top\mathbf D\mathbf XX⊤DX invertible. The book substitutes a pseudoinverse otherwise; that case is not formalized, and without the hypothesis Lean's matrix inverse is zero and the PBE is trivially 000.
  • Coverage (π(a∣s)>0⇒b(a∣s)>0\pi(a\mid s)>0\Rightarrow b(a\mid s)>0π(a∣s)>0⇒b(a∣s)>0) for the expectation identities.
  • 0≤γ<10\le\gamma<10≤γ<1 for (11.21); the episodic γ=1\gamma=1γ=1 case is not stated.
  • In (11.24) the return enters through conditional distributions νs\nu_sνs​ with finite second moment and mean vπ(s)v_\pi(s)vπ​(s).

Readings and corrections:

  • (11.10) is introduced as minimizing "the VE", but its displayed objective is an unweighted sum over the two states. The displayed sum is formalized.
  • The sentence on p. 269, "there always exists an approximate value function with zero PBE", needs X⊤D(I−γPπ)X\mathbf X^\top\mathbf D(I-\gamma P_\pi)\mathbf XX⊤D(I−γPπ​)X invertible, which can fail off-policy. Counterexample: two states with features x=1,2x = 1, 2x=1,2, both moving to the second state under π\piπ, rewards rπ=(1,0)r_\pi = (1, 0)rπ​=(1,0), γ=34\gamma = \tfrac34γ=43​ and μ=(23,13)\mu = (\tfrac23, \tfrac13)μ=(32​,31​). Then X⊤Dδˉw=23\mathbf X^\top\mathbf D\bar\delta_{\mathbf w} = \tfrac23X⊤Dδˉw​=32​ for every w\mathbf ww, so PBE(w)=(23)2/2>0\mathrm{PBE}(\mathbf w) = (\tfrac23)^2/2 > 0PBE(w)=(32​)2/2>0 everywhere. The sentence is not formalized.
  • The Example 11.4 MRPs are transcribed from the figure on p. 276. Their equal data distributions are stated through all finite prefixes of the observed stream (feature vectors and rewards) from the stationary distributions; the BE minimizer of the second MRP as γ→1\gamma\to1γ→1 is not formalized.
  • Baird's counterexample (divergence shown by simulation), Gradient-TD and Emphatic-TD convergence (asserted with references) are out of scope.

A trivializing formalization is ruled out: vπv_\pivπ​ is not defined as a fixed point of BπB_\piBπ​, the PBE is not stated without the invertibility hypothesis, and the gradient is a derivative of the defined PBE, not a restated formula.

Welcome contributions: proofs of the milestones, and a general lemma on the gradient of h⊤C−1h\mathbf h^\top\mathbf C^{-1}\mathbf hh⊤C−1h for affine h\mathbf hh and symmetric invertible C\mathbf CC, which is reusable well beyond this mission.

Selected references

  • R. S. Sutton and A. G. Barto, Reinforcement Learning: An Introduction, 2nd ed., MIT Press, 2018, ISBN 9780262039246, Chapter 11. http://incompleteideas.net/book/the-book-2nd.html
  • L. Baird, Residual algorithms: Reinforcement learning with function approximation, ICML 1995. https://doi.org/10.1016/B978-1-55860-377-6.50013-X
  • J. N. Tsitsiklis and B. Van Roy, Feature-based methods for large scale dynamic programming, Machine Learning 22, 1996. https://doi.org/10.1007/BF00114724
  • R. S. Sutton, H. R. Maei, D. Precup, S. Bhatnagar, D. Silver, Cs. Szepesvári, E. Wiewiora, Fast gradient-descent methods for temporal-difference learning with linear function approximation, ICML 2009. https://doi.org/10.1145/1553374.1553501
  • R. S. Sutton, A. R. Mahmood, M. White, An emphatic approach to the problem of off-policy temporal-difference learning, JMLR 17, 2016. https://jmlr.org/papers/v17/14-488.html
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Algorithmic Game TheoryConvex OptimizationOptimization·Captain: mikedeng1

Blackwell Approachability and No-Regret Learning are Equivalent 3: An Efficient Forecaster Whose (ℓ1, ε)-Calibration Rate Is at Most √(2/(εT))Research Paper

Calibrated forecasting

A forecaster announces, each day, a probability that it will rain; afterwards nature reveals whether it did. The forecaster is calibrated if, on the days on which it announced roughly 30%, it rained roughly 30% of the time, and likewise for every other announced value. Calibration is a minimal consistency requirement for probabilistic forecasts, used in meteorology, in the evaluation of probabilistic classifiers, and in game theory, where calibrated forecasts of the opponents' play lead to correlated equilibrium (Foster and Vohra, 1997).

Calibration is achievable even against an adversary who chooses the outcomes, provided the forecaster randomizes. Timeline:

  • 1998. Foster and Vohra construct an asymptotically calibrated randomized forecaster against an arbitrary outcome sequence.
  • 1999. Foster reduces calibration to Blackwell's approachability theorem by exhibiting, for each halfspace, a forecast that keeps the payoff inside it.
  • 2009. Mannor and Stoltz give an approachability-based calibration procedure concurrently with the paper below.
  • 2011. Abernethy, Bartlett and Hazan prove that Blackwell approachability and no-regret online linear optimization are equivalent, and use the equivalence to obtain an efficient calibrated forecaster: O(log⁡1/ε)O(\log 1/\varepsilon)O(log1/ε) time per round and calibration rate O(1/εT)O(1/\sqrt{\varepsilon T})O(1/εT​).

This mission formalizes the last result, Theorem 22 of the 2011 paper, in the explicit form given by its proof.

Setting

Fix a positive integer mmm and the grid width ε=1/m\varepsilon = 1/mε=1/m. Each round t=1,…,Tt = 1, \dots, Tt=1,…,T the forecaster chooses a probability vector wtw_twt​ in the simplex Δm+1\Delta_{m+1}Δm+1​ over the grid indices i=0,…,mi = 0, \dots, mi=0,…,m, draws it∼wti_t \sim w_tit​∼wt​ and announces pt=it/mp_t = i_t/mpt​=it​/m. Nature then reveals yt∈{0,1}y_t \in \{0, 1\}yt​∈{0,1}.

Vectors live in Rm+1\mathbb R^{m+1}Rm+1 with the Euclidean norm ∥⋅∥2\|\cdot\|_2∥⋅∥2​. The ℓ₁ norm is ∥x∥1=∑i∣xi∣\|x\|_1 = \sum_i |x_i|∥x∥1​=∑i​∣xi​∣, the ℓ₁ ball is B1(r)={y:∥y∥1≤r}B_1(r) = \{y : \|y\|_1 \le r\}B1​(r)={y:∥y∥1​≤r}, and the unit cube is B∞(1)={θ:∣θi∣≤1 for all i}B_\infty(1) = \{\theta : |\theta_i| \le 1 \text{ for all } i\}B∞​(1)={θ:∣θi​∣≤1 for all i}.

The calibration game (11) has payoff

u(w,y)=(w(0)(y−0m), w(1)(y−1m), …, w(m)(y−1))∈Rm+1.u(w, y) = \Bigl(w(0)\bigl(y - \tfrac0m\bigr),\ w(1)\bigl(y - \tfrac1m\bigr),\ \dots,\ w(m)(y - 1)\Bigr) \in \mathbb R^{m+1}.u(w,y)=(w(0)(y−m0​), w(1)(y−m1​), …, w(m)(y−1))∈Rm+1.

The (ℓ1,ε)(\ell_1, \varepsilon)(ℓ1​,ε)-calibration rate (Definition 19) of the announced forecasts is max⁡{0,∑i=0m∣1T∑t=1TI[pt=i/m](i/m−yt)∣−ε/2}\max\{0, \sum_{i=0}^m |\frac1T\sum_{t=1}^T \mathbb I[p_t = i/m](i/m - y_t)| - \varepsilon/2\}max{0,∑i=0m​∣T1​∑t=1T​I[pt​=i/m](i/m−yt​)∣−ε/2}. Replacing each indicator by its expectation wt(i)w_t(i)wt​(i) gives the rate of the forecast distributions,

CˉTε=max⁡{0, ∑i=0m∣1T∑t=1Twt(i)(im−yt)∣−ε2},\bar C^\varepsilon_T = \max\Bigl\{0,\ \sum_{i=0}^m \Bigl|\frac1T \sum_{t=1}^T w_t(i)\Bigl(\frac im - y_t\Bigr)\Bigr| - \frac\varepsilon2\Bigr\},CˉTε​=max{0, i=0∑m​​T1​t=1∑T​wt​(i)(mi​−yt​)​−2ε​},

which is max⁡{0,∥uˉT∥1−ε/2}\max\{0, \|\bar u_T\|_1 - \varepsilon/2\}max{0,∥uˉT​∥1​−ε/2} for the average payoff uˉT=1T∑tu(wt,yt)\bar u_T = \frac1T\sum_t u(w_t, y_t)uˉT​=T1​∑t​u(wt​,yt​).

The forecaster is Algorithm 5. It keeps a point θt\theta_tθt​ in the cube, starting from θ1=0\theta_1 = 0θ1​=0 with w1w_1w1​ arbitrary. After round ttt it takes a projected gradient step (Algorithm 4, online gradient descent) against the loss vector ft=−u(wt,yt)f_t = -u(w_t, y_t)ft​=−u(wt​,yt​):

θt+1=ΠB∞(1)(θt+η u(wt,yt)),\theta_{t+1} = \Pi_{B_\infty(1)}\bigl(\theta_t + \eta\, u(w_t, y_t)\bigr),θt+1​=ΠB∞​(1)​(θt​+ηu(wt​,yt​)),

where Π\PiΠ is the Euclidean projection. It then sets wt+1w_{t+1}wt+1​ to the output of the oracle Algorithm 3 on θt+1\theta_{t+1}θt+1​, which puts weight on at most two adjacent grid points where θ\thetaθ changes sign.

Formalization targets

Goal: Theorem 22 in the form (14)

For m≥1m \ge 1m≥1, T≥1T \ge 1T≥1, every outcome sequence y1,…,yT∈{0,1}y_1, \dots, y_T \in \{0, 1\}y1​,…,yT​∈{0,1} and every run of Algorithm 5 with η=(m+1)/T\eta = \sqrt{(m+1)/T}η=(m+1)/T​,

CˉTε≤2εT.\bar C^\varepsilon_T \le \sqrt{\frac{2}{\varepsilon T}}.CˉTε​≤εT2​​.

This is the bound CTε≤GD/TC^\varepsilon_T \le GD/\sqrt TCTε​≤GD/T​ of display (14) with the paper's constant G=2G = \sqrt 2G=2​.

Milestones

  1. Claim 1 (proof): min⁡∥y∥1≤ε/2∥x−y∥1=max⁡{0,−ε/2+∥x∥1}\min_{\|y\|_1 \le \varepsilon/2}\|x - y\|_1 = \max\{0, -\varepsilon/2 + \|x\|_1\}min∥y∥1​≤ε/2​∥x−y∥1​=max{0,−ε/2+∥x∥1​}.
  2. Display (13): for ∥x∥1>ε/2\|x\|_1 > \varepsilon/2∥x∥1​>ε/2, also =−ε/2−min⁡∥θ∥∞≤1⟨−x,θ⟩= -\varepsilon/2 - \min_{\|\theta\|_\infty \le 1}\langle -x, \theta\rangle=−ε/2−min∥θ∥∞​≤1​⟨−x,θ⟩.
  3. Algorithm 3: for every θ\thetaθ in the cube there is an output w∈Δm+1w \in \Delta_{m+1}w∈Δm+1​, and every output satisfies ⟨u(w,y),θ⟩≤ε/2\langle u(w, y), \theta\rangle \le \varepsilon/2⟨u(w,y),θ⟩≤ε/2 for all y∈[0,1]y \in [0, 1]y∈[0,1].
  4. Display (12): under that guarantee, max⁡{0,∥uˉT∥1−ε/2}≤1T(∑t⟨−ut,θt⟩−min⁡θ∈B∞(1)∑t⟨−ut,θ⟩)\max\{0, \|\bar u_T\|_1 - \varepsilon/2\} \le \frac1T\bigl(\sum_t \langle -u_t, \theta_t\rangle - \min_{\theta \in B_\infty(1)}\sum_t\langle -u_t, \theta\rangle\bigr)max{0,∥uˉT​∥1​−ε/2}≤T1​(∑t​⟨−ut​,θt​⟩−minθ∈B∞​(1)​∑t​⟨−ut​,θ⟩).
  5. Online gradient descent: regret at most DGTDG\sqrt TDGT​ with step η=D/(GT)\eta = D/(G\sqrt T)η=D/(GT​).
  6. Theorem 21 (response-satisfiability and approachability): for every y∈[0,1]y \in [0,1]y∈[0,1] some w∈Δm+1w \in \Delta_{m+1}w∈Δm+1​ has u(w,y)∈B1(ε/2)u(w, y) \in B_1(\varepsilon/2)u(w,y)∈B1​(ε/2); hence some algorithm choosing wtw_twt​ from y1,…,yt−1y_1, \dots, y_{t-1}y1​,…,yt−1​ drives the distance of the average payoff to B1(ε/2)B_1(\varepsilon/2)B1​(ε/2) to 000 against every outcome sequence in [0,1][0,1][0,1].

Significance

The bound shows that a forecaster with logarithmic per-round cost has calibration error vanishing at rate T−1/2T^{-1/2}T−1/2 against every outcome sequence. Earlier calibrated forecasters required solving a linear program or computing a fixed point each round. The construction is also the paper's worked instance of its general equivalence: a calibration problem, posed as approachability of an ℓ₁ ball, is solved by a no-regret learner on the dual unit cube together with a halfspace oracle.

The result is proved in the paper; no machine-checked version is known to exist. The formalization makes explicit three points the paper leaves informal: the step size, the sign of the gradient step, and the gap between the forecast distributions and the sampled forecasts. The milestones are reusable on their own: the ℓ₁/ℓ∞ duality, and the regret bound of online gradient descent for linear losses on a general closed convex set.

Difficulty

The chain (12)–(14) looks like a direct composition, but each link has content. The oracle guarantee needs a case analysis over the sign pattern of θ\thetaθ, including the degenerate case θ(i+1)=0\theta(i+1) = 0θ(i+1)=0. The reduction (12) needs the duality (13) with attained minima, and it holds only outside the ball B1(ε/2)B_1(\varepsilon/2)B1​(ε/2). The regret bound of online gradient descent needs the non-expansiveness of the Euclidean projection and a telescoping argument. The tempting shortcut of quoting "OGD has regret O(T)O(\sqrt T)O(T​)" does not give the stated constant without fixing the step size.

Formalization scope

Vectors are EuclideanSpace ℝ (Fin (m+1)), with grid index i∈{0,…,m}i \in \{0, \dots, m\}i∈{0,…,m} as Fin (m+1) and i/mi/mi/m as a real quotient; the ℓ₁ norm and the cube are written out coordinatewise. Rounds are t=1,…,Tt = 1, \dots, Tt=1,…,T. Minima over sets are stated through IsLeast or as the infimum of the image of a nonempty bounded set. Algorithm 3 is a relation that allows every sign-change index the binary search might return. The projection is any Euclidean minimizer onto the cube.

Conventions and corrections, each disclosed in the item's Formalization Note:

  • Gradient-step sign. Algorithm 4 prints θt−ηut\theta_t - \eta u_tθt​−ηut​, but the proof runs the learner on the losses ft=−utf_t = -u_tft​=−ut​ (condition 2), so the step is θt+ηut\theta_t + \eta u_tθt​+ηut​. With the printed sign the bound fails.
  • Step size. The page sets η=O(T−1/2)\eta = O(T^{-1/2})η=O(T−1/2); the goal pins η=(m+1)/T\eta = \sqrt{(m+1)/T}η=(m+1)/T​, the standard tuning with radius m+1\sqrt{m+1}m+1​ of the cube and ∥ut∥2≤1\|u_t\|_2 \le 1∥ut​∥2​≤1. The page's D=1/εD = \sqrt{1/\varepsilon}D=1/ε​ is not the cube's diameter.
  • Forecast distributions. The rate is that of the distributions wtw_twt​, the expectation of the calibration vector over the forecaster's draws (Lemma 20). The high-probability statement for the sampled forecasts is not formalized, nor is the running-time claim.
  • Other misprints. Algorithm 3's header "w↦θw \mapsto \thetaw↦θ" is θ↦w\theta \mapsto wθ↦w, and the calibration vector has m+1m + 1m+1 coordinates, not ⌊ε−1⌋\lfloor \varepsilon^{-1} \rfloor⌊ε−1⌋.
  • Added hypotheses. m≥1m \ge 1m≥1 and T≥1T \ge 1T≥1.

A trivializing formalization is ruled out. The rate is defined from Definition 19's formula, not as a distance, and the step size is pinned. A free step size would make the bound false, and an empty oracle relation would make it vacuous; milestone 3's existence clause excludes the latter.

Contributions are welcome on each milestone. The online gradient descent bound and the ℓ₁/ℓ∞ duality are independent of calibration. The published one-step inequality LogRegretOCO.OGD.one_step_inequality is included as a reference item for the regret bound.

Selected references

  • J. Abernethy, P. L. Bartlett, E. Hazan, Blackwell Approachability and No-Regret Learning are Equivalent, COLT 2011, JMLR W&CP 19, pp. 27–46, 2011. https://proceedings.mlr.press/v19/abernethy11b.html
  • D. P. Foster, R. V. Vohra, Asymptotic calibration, Biometrika 85(2), 1998. https://doi.org/10.1093/biomet/85.2.379
  • D. P. Foster, A proof of calibration via Blackwell's approachability theorem, Games and Economic Behavior 29, 1999. https://doi.org/10.1006/game.1999.0724
  • D. P. Foster, R. V. Vohra, Calibrated learning and correlated equilibrium, Games and Economic Behavior 21, 1997. https://doi.org/10.1006/game.1997.0595
  • S. Mannor, G. Stoltz, A geometric proof of calibration, Mathematics of Operations Research 35(4), 2010. https://arxiv.org/abs/0908.3576
  • M. Zinkevich, Online convex programming and generalized infinitesimal gradient ascent, ICML 2003. https://dl.acm.org/doi/10.5555/3041838.3041955
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ProbabilityReinforcement LearningStatistics·Captain: mikedeng1

Reinforcement Learning: An Introduction V: Off-policy Prediction by Importance SamplingTextbook

Motivation

Reinforcement learning methods must explore in order to find good behaviour, yet the quantity they usually want to evaluate is the value of a different, often deterministic, policy. Off-policy prediction separates the two roles: episodes are generated by a behaviour policy bbb, and the goal is the value function vπv_\pivπ​ of a target policy π\piπ. Almost every off-policy method in Sutton and Barto's Reinforcement Learning: An Introduction (2nd ed., MIT Press, 2018), and in the literature that follows it, rests on importance sampling: a return observed under bbb is reweighted by the relative probability of its trajectory under π\piπ and bbb. Section 5.5 of the book introduces the idea for Monte Carlo prediction, §5.6 gives the incremental form of the weighted estimator, and §§5.8–5.9 refine the weights using the internal structure of the return: discounting-aware importance sampling, after Sutton, Mahmood, Precup and van Hasselt (2014), and per-decision importance sampling, introduced by Precup, Sutton and Singh (2000). The book's remarks on the variance of the two estimators (p. 105) cite Precup, Sutton and Dasgupta (2001). Later chapters (7, 11, 12) reuse the same ratios for nnn-step, gradient-TD and eligibility-trace methods.

This mission is the fifth in a series formalizing the book's central mathematical claims. It covers §§5.5–5.9 (pp. 103–115).

Setting

A finite Markov decision process has finite sets of states S\mathcal SS (terminal states included), actions A\mathcal AA and rewards R⊂R\mathcal R \subset \mathbb RR⊂R, and dynamics p(s′,r∣s,a)p(s', r \mid s, a)p(s′,r∣s,a): for each (s,a)(s, a)(s,a) a probability distribution over next state and reward. The state-transition probability is p(s′∣s,a)=∑rp(s′,r∣s,a)p(s' \mid s, a) = \sum_r p(s', r \mid s, a)p(s′∣s,a)=∑r​p(s′,r∣s,a). A policy μ\muμ gives a distribution μ(⋅∣s)\mu(\cdot \mid s)μ(⋅∣s) over actions in every state.

An episode from a start state sss is a sequence S0=s,A0,R1,S1,…,AT−1,RT,STS_0 = s, A_0, R_1, S_1, \dots, A_{T-1}, R_T, S_TS0​=s,A0​,R1​,S1​,…,AT−1​,RT​,ST​ in which S0,…,ST−1S_0, \dots, S_{T-1}S0​,…,ST−1​ are nonterminal and STS_TST​ is terminal. Under μ\muμ it has probability ∏k=0T−1μ(Ak∣Sk) p(Sk+1,Rk+1∣Sk,Ak)\prod_{k=0}^{T-1} \mu(A_k \mid S_k)\, p(S_{k+1}, R_{k+1} \mid S_k, A_k)∏k=0T−1​μ(Ak​∣Sk​)p(Sk+1​,Rk+1​∣Sk​,Ak​). The return is G0=∑k=0T−1γkRk+1G_0 = \sum_{k=0}^{T-1} \gamma^k R_{k+1}G0​=∑k=0T−1​γkRk+1​ with discount rate γ∈[0,1]\gamma \in [0, 1]γ∈[0,1], and the value vπ(s)v_\pi(s)vπ​(s) is the expected return of an episode generated by π\piπ from sss.

The behaviour policy covers the target policy if π(a∣s)>0\pi(a \mid s) > 0π(a∣s)>0 implies b(a∣s)>0b(a \mid s) > 0b(a∣s)>0. The importance-sampling ratio of decisions 0,…,j0, \dots, j0,…,j is

ρ0:j=∏k=0jπ(Ak∣Sk)b(Ak∣Sk),\rho_{0:j} = \prod_{k=0}^{j} \frac{\pi(A_k \mid S_k)}{b(A_k \mid S_k)},ρ0:j​=k=0∏j​b(Ak​∣Sk​)π(Ak​∣Sk​)​,

and the per-decision return weights each reward only by the ratio of the decisions that precede it:

G~0=ρ0:0R1+γρ0:1R2+⋯+γT−1ρ0:T−1RT.\tilde G_0 = \rho_{0:0} R_1 + \gamma \rho_{0:1} R_2 + \dots + \gamma^{T-1} \rho_{0:T-1} R_T .G~0​=ρ0:0​R1​+γρ0:1​R2​+⋯+γT−1ρ0:T−1​RT​.

The book writes these objects at a general time ttt and conditions on St=sS_t = sSt​=s; by the Markov property this is the same as starting the episode at sss, which is what the formal statements do.

Formalization targets

Goal: unbiasedness of ordinary and per-decision importance sampling

For episodes generated by bbb from sss,

Eb[ρ0:T−1G0∣S0=s]=vπ(s)=Eb[G~0∣S0=s].\mathbb E_b\bigl[\rho_{0:T-1} G_0 \mid S_0 = s\bigr] = v_\pi(s) = \mathbb E_b\bigl[\tilde G_0 \mid S_0 = s\bigr].Eb​[ρ0:T−1​G0​∣S0​=s]=vπ​(s)=Eb​[G~0​∣S0​=s].

The first equality is Eq. (5.4) (p. 104); the second is the statement E[ρt:T−1Gt]=E[G~t]\mathbb E[\rho_{t:T-1}G_t] = \mathbb E[\tilde G_t]E[ρt:T−1​Gt​]=E[G~t​] of §5.9 (p. 114).

Milestones

  1. (5.3): the trajectory probability is a product, and the ratio of trajectory probabilities under π\piπ and bbb is ρ0:T−1\rho_{0:T-1}ρ0:T−1​, independent of the dynamics.
  2. (5.4) alone.
  3. (5.13): ∑ab(a∣x) π(a∣x)/b(a∣x)=∑aπ(a∣x)=1\sum_a b(a \mid x)\, \pi(a \mid x)/b(a \mid x) = \sum_a \pi(a \mid x) = 1∑a​b(a∣x)π(a∣x)/b(a∣x)=∑a​π(a∣x)=1 under coverage.
  4. (5.14) and its kkk-th form: Eb[ρ0:T−1Rk]=Eb[ρ0:k−1Rk]\mathbb E_b[\rho_{0:T-1} R_k] = \mathbb E_b[\rho_{0:k-1} R_k]Eb​[ρ0:T−1​Rk​]=Eb​[ρ0:k−1​Rk​] for every k≥1k \ge 1k≥1 (Exercise 5.13).
  5. Example 5.5: in a one-state MDP with a loop, vπ(s)=1v_\pi(s) = 1vπ​(s)=1 and Eb[ρ0:T−1G0]=1\mathbb E_b[\rho_{0:T-1}G_0] = 1Eb​[ρ0:T−1​G0​]=1, yet Eb[(ρ0:T−1G0)2]=∞\mathbb E_b[(\rho_{0:T-1}G_0)^2] = \inftyEb​[(ρ0:T−1​G0​)2]=∞.
  6. (5.7)–(5.8): the incremental rule Vn+1=Vn+(Wn/Cn)(Gn−Vn)V_{n+1} = V_n + (W_n/C_n)(G_n - V_n)Vn+1​=Vn​+(Wn​/Cn​)(Gn​−Vn​) computes the weighted average ∑k<nWkGk/∑k<nWk\sum_{k<n} W_k G_k / \sum_{k<n} W_k∑k<n​Wk​Gk​/∑k<n​Wk​ (Exercise 5.10).
  7. §5.8: Gt=(1−γ)∑h=t+1T−1γh−t−1Gˉt:h+γT−t−1Gˉt:TG_t = (1-\gamma)\sum_{h=t+1}^{T-1}\gamma^{h-t-1}\bar G_{t:h} + \gamma^{T-t-1}\bar G_{t:T}Gt​=(1−γ)∑h=t+1T−1​γh−t−1Gˉt:h​+γT−t−1Gˉt:T​ with flat partial returns Gˉt:h=Rt+1+⋯+Rh\bar G_{t:h} = R_{t+1} + \dots + R_hGˉt:h​=Rt+1​+⋯+Rh​.

Significance

Eq. (5.4) is the reason the first-visit ordinary importance-sampling estimator (5.5) is unbiased, and it is the template for every importance-sampling correction in the rest of the book. The per-decision identity shows that an estimator with fewer ratio factors per reward, (5.15), has the same expectation, which is the starting point for per-decision and control-variate methods for multi-step off-policy learning (Precup, Sutton and Singh 2000). Example 5.5 shows that unbiasedness says nothing about variance: the ordinary estimator can have infinite variance on a two-action problem, which motivates weighted importance sampling and the incremental weighted update of §5.6.

The results of these sections are classical and proved informally in the book, partly as exercises (5.10, 5.13) left without solution. No machine-checked version exists on the platform: a search for importance sampling, off-policy and per-decision returned no statements. The mission produces a formal trajectory model of an episodic MDP under two policies, which later missions on nnn-step off-policy returns and off-policy traces can reuse.

Difficulty

Eq. (5.4) itself is a termwise identity: for every episode, Pr⁡b(episode) ρ0:T−1=Pr⁡π(episode)\Pr_b(\text{episode})\,\rho_{0:T-1} = \Pr_\pi(\text{episode})Prb​(episode)ρ0:T−1​=Prπ​(episode) under coverage. The per-decision identity is not termwise. The later factors of ρ0:T−1\rho_{0:T-1}ρ0:T−1​ multiply a reward that was received before the corresponding decisions, and removing them requires summing over all continuations of an episode prefix, of every remaining length, and using that each factor has conditional expectation one (5.13) and that the continuation terminates with probability one. The obvious attempt, cancelling the factors episode by episode, fails: on a single episode ρ0:T−1R1\rho_{0:T-1}R_1ρ0:T−1​R1​ and ρ0:0R1\rho_{0:0}R_1ρ0:0​R1​ differ.

In Example 5.5 the episodes have no length bound, so the expected square is an infinite series over episode lengths whose divergence must be shown directly.

Formalization scope

  • States, actions and rewards are finite types; the terminal states are a finite subset of the state type. Policies are stochastic, one action set is used in every state, and vπv_\pivπ​ is defined as an expected return, never as the solution of a Bellman equation.
  • Expectations are series over episode lengths of finite sums over episodes. Lean assigns 000 to a divergent series, so the goal and milestones 2 and 4 assume that under bbb every episode from sss terminates within a fixed number HHH of steps with probability one. The book leaves termination implicit; this bounded-horizon hypothesis is a restriction relative to the book's episodic setting and is stated as such. Example 5.5, whose episodes are unbounded, is stated without it, with the expected square in [0,∞][0, \infty][0,∞].
  • The discount rate is kept general in [0,1][0, 1][0,1].
  • The importance-sampling ratio uses real division; a factor with b(Ak∣Sk)=0b(A_k \mid S_k) = 0b(Ak​∣Sk​)=0 evaluates to 000 in Lean, but such episodes have probability 000 under bbb.
  • The flat-partial-return decomposition is an algebraic identity and is stated for every real γ\gammaγ, which is more general than the book's "for any γ∈[0,1)\gamma \in [0,1)γ∈[0,1)".
  • The incremental weighted update is stated with nonnegative weights and W1>0W_1 > 0W1​>0. The book's C0=0C_0 = 0C0​=0 makes (5.8) divide by zero at n=1n = 1n=1 when W1=0W_1 = 0W1​=0; the hypothesis excludes that case.
  • A trivializing formalization is ruled out: vπv_\pivπ​ is the expected return of π\piπ's own episodes, the ratio is computed from the episode, and the per-decision identity, which carries the chapter's content beyond (5.4), is part of the goal.
  • Not stated: the bias and variance comparisons of ordinary and weighted importance sampling (p. 105) and the discounting-aware estimators (5.9)–(5.10) as estimators; these are statistical claims about estimators over a random number of visits that the book does not make precise.

Contributions welcome: proofs of the milestones, a general measure-theoretic version of the trajectory model without the bounded-horizon hypothesis, and variants for action values qπq_\piqπ​ (Exercise 5.6).

Selected references

  • R. S. Sutton and A. G. Barto, Reinforcement Learning: An Introduction, 2nd ed., MIT Press, 2018, ISBN 9780262039246, §§5.5–5.9, pp. 103–115. http://incompleteideas.net/book/the-book-2nd.html
  • D. Precup, R. S. Sutton and S. Singh, Eligibility Traces for Off-Policy Policy Evaluation, Proceedings of the 17th International Conference on Machine Learning (ICML), 2000, pp. 759–766 (cited in the book's bibliography).
  • D. Precup, R. S. Sutton and S. Dasgupta, Off-Policy Temporal-Difference Learning with Function Approximation, Proceedings of the 18th International Conference on Machine Learning (ICML), 2001, pp. 417–424 (cited in the book, p. 105).
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Analysis of Thompson Sampling for the Multi-armed Bandit Problem 1: Logarithmic Regret for Two ArmsResearch Paper

Motivation

Thompson Sampling (TS) is the oldest heuristic for the stochastic multi-armed bandit problem: it was proposed by Thompson in 1933 (Biometrika 25) and is used in practice for online advertising and recommendation, where it often performs as well as or better than upper-confidence-bound methods (Chapelle and Li, NIPS 2011; Scott 2010). Until 2012 its theoretical guarantees for the frequentist regret were weak: earlier analyses gave only o(T)o(T)o(T) regret in time TTT (Granmo 2010; May, Korda, Lee and Leslie 2011).

Agrawal and Goyal, Analysis of Thompson Sampling for the Multi-armed Bandit Problem (arXiv:1111.1797v3, COLT 2012), gave the first logarithmic finite-time bound on the expected regret of TS. This mission formalizes their two-armed result, Theorem 1. A companion mission covers the NNN-armed bound, Theorem 2.

Timeline. Lai and Robbins (1985) proved that every consistent algorithm has regret at least [∑iΔi/D(μi∥μ∗)+o(1)]ln⁡T\big[\sum_i \Delta_i/D(\mu_i\|\mu^*)+o(1)\big]\ln T[∑i​Δi​/D(μi​∥μ∗)+o(1)]lnT. Auer, Cesa-Bianchi and Fischer (2002) gave UCB1 with an O(∑iln⁡T/Δi)O(\sum_i \ln T/\Delta_i)O(∑i​lnT/Δi​) finite-time bound. Agrawal and Goyal (2012) proved O(ln⁡T/Δ+1/Δ3)O(\ln T/\Delta+1/\Delta^3)O(lnT/Δ+1/Δ3) for two-armed TS. Kaufmann, Korda and Munos (ALT 2012) and Agrawal and Goyal (AISTATS 2013) later proved asymptotically optimal bounds for Bernoulli TS.

Setting

There are two arms. Arm i∈{1,2}i\in\{1,2\}i∈{1,2} has a fixed, unknown reward distribution DiD_iDi​ supported in [0,1][0,1][0,1], with mean μi\mu_iμi​. Plays of an arm give i.i.d. rewards, independent of the other arm. Arm 1 is the unique optimal arm, μ1>μ2\mu_1>\mu_2μ1​>μ2​, and Δ=μ1−μ2\Delta=\mu_1-\mu_2Δ=μ1​−μ2​ is the gap.

Thompson Sampling for general stochastic bandits (Algorithm 2 of the paper) keeps, for each arm iii, a success count SiS_iSi​ and a failure count FiF_iFi​, both starting at 000. In each round t=1,2,…t=1,2,\dotst=1,2,… it

  1. samples, independently for each arm, θi(t)∼Beta(Si+1,Fi+1)\theta_i(t)\sim\mathrm{Beta}(S_i+1,F_i+1)θi​(t)∼Beta(Si​+1,Fi​+1);
  2. plays i(t)=arg⁡max⁡iθi(t)i(t)=\arg\max_i\theta_i(t)i(t)=argmaxi​θi​(t) and observes a reward r~t∼Di(t)\tilde r_t\sim D_{i(t)}r~t​∼Di(t)​;
  3. performs a Bernoulli trial with success probability r~t\tilde r_tr~t​, with outcome rt∈{0,1}r_t\in\{0,1\}rt​∈{0,1};
  4. increments Si(t)S_{i(t)}Si(t)​ if rt=1r_t=1rt​=1 and Fi(t)F_{i(t)}Fi(t)​ otherwise.

ki(t)k_i(t)ki​(t) is the number of plays of arm iii before round ttt. The expected regret in time TTT is

E[R(T)]=E[∑t=1T(μ1−μi(t))],\mathbb E[\mathcal R(T)]=\mathbb E\Big[\sum_{t=1}^T(\mu_1-\mu_{i(t)})\Big],E[R(T)]=E[t=1∑T​(μ1​−μi(t)​)],

the expectation being over the rewards and the algorithm's randomness.

The analysis uses the Beta cdf Fα,βbetaF^{beta}_{\alpha,\beta}Fα,βbeta​, the binomial cdf Fn,pBF^B_{n,p}Fn,pB​, and the random variable X(j,s,y)X(j,s,y)X(j,s,y): the number of independent Beta(s+1,j−s+1)\mathrm{Beta}(s+1,j-s+1)Beta(s+1,j−s+1) draws made before one exceeds yyy.

Formalization targets

Goal: Theorem 1 (p. 3)

There is an absolute constant C>0C>0C>0 such that for every two-armed instance with rewards in [0,1][0,1][0,1] and μ1>μ2\mu_1>\mu_2μ1​>μ2​, and every T≥2T\ge 2T≥2,

E[R(T)]≤C(ln⁡TΔ+1Δ3).\mathbb E[\mathcal R(T)]\le C\Big(\frac{\ln T}{\Delta}+\frac1{\Delta^3}\Big).E[R(T)]≤C(ΔlnT​+Δ31​).

The constant is not fixed numerically: the paper states the theorem in O(⋅)O(\cdot)O(⋅) form (footnote 1), and the explicit display it reports on p. 8, 40ln⁡T/Δ+48/Δ3+18Δ40\ln T/\Delta+48/\Delta^3+18\Delta40lnT/Δ+48/Δ3+18Δ, is not the formal claim.

Milestones

  • Fact 1 (p. 12): Fα,βbeta(y)=1−Fα+β−1,yB(α−1)F^{beta}_{\alpha,\beta}(y)=1-F^B_{\alpha+\beta-1,y}(\alpha-1)Fα,βbeta​(y)=1−Fα+β−1,yB​(α−1) for positive integers α,β\alpha,\betaα,β.
  • Lemma 1 (p. 6): E[X(j,s,y)]=1/Fj+1,yB(s)−1\mathbb E[X(j,s,y)]=1/F^B_{j+1,y}(s)-1E[X(j,s,y)]=1/Fj+1,yB​(s)−1.
  • Lemma 6 (p. 13): Hoeffding-type bounds (10)–(11) on binomial cdfs.
  • Fact 2 (p. 13): every median of Binomial(n,p)\mathrm{Binomial}(n,p)Binomial(n,p) is ⌊np⌋\lfloor np\rfloor⌊np⌋ or ⌈np⌉\lceil np\rceil⌈np⌉.
  • Lemma 2 (p. 7): Pr⁡(E2(t))≥1−2/T2\Pr(E_2(t))\ge 1-2/T^2Pr(E2​(t))≥1−2/T2, where E2(t)={θ2(t)≤μ2+Δ/2 or k2(t)<24ln⁡T/Δ2}E_2(t)=\{\theta_2(t)\le\mu_2+\Delta/2\ \text{or}\ k_2(t)<24\ln T/\Delta^2\}E2​(t)={θ2​(t)≤μ2​+Δ/2 or k2​(t)<24lnT/Δ2}.
  • Lemma 3 (p. 7): a three-case bound on E[E[min⁡{X(j,s(j),y),T}∣s(j)]]\mathbb E\big[\mathbb E[\min\{X(j,s(j),y),T\}\mid s(j)]\big]E[E[min{X(j,s(j),y),T}∣s(j)]] for s(j)∼Binomial(j,μ1)s(j)\sim\mathrm{Binomial}(j,\mu_1)s(j)∼Binomial(j,μ1​).
  • Eq. (1) (p. 7): E[k2(T)]≤C(ln⁡T/Δ2+1/Δ4)\mathbb E[k_2(T)]\le C(\ln T/\Delta^2+1/\Delta^4)E[k2​(T)]≤C(lnT/Δ2+1/Δ4).

Significance

The result. Theorem 1 shows that TS, a randomized Bayesian heuristic with no explicit confidence bonus, has regret logarithmic in TTT on every two-armed instance, matching the order in TTT of the Lai–Robbins lower bound. The proof introduced a way to control the optimal arm's waiting time between plays through the Beta–Binomial duality, and later analyses of TS reuse that device.

Formalizing it. The result is proved on paper and has no machine-checked proof that we know of. The platform's existing TS results concern Gaussian TS (Lattimore and Szepesvári, Ch. 36) and Bayesian regret, which are different algorithms or regret notions. A formalization adds a reusable Lean model of Algorithm 2 on [0,1][0,1][0,1]-valued rewards, Beta–Binomial facts (Fact 1, Lemma 1), a binomial-median theorem, and binomial Hoeffding bounds. It also produces a proof with a constant that has been checked, since the printed constants contain an arithmetic slip.

Difficulty

The standard UCB argument does not transfer to TS. For UCB, the optimal arm's index exceeds its mean with high probability however often the arm has been played, because the exploration bonus is deterministic; the analysis then only has to count plays of the suboptimal arm until its own index concentrates, after Θ(ln⁡T/Δ2)\Theta(\ln T/\Delta^2)Θ(lnT/Δ2) plays. Under TS the optimal arm's sample θ1(t)\theta_1(t)θ1​(t) is random and, if the arm has been played rarely or its early rewards were poor, it falls below μ2\mu_2μ2​ with constant probability. The optimal arm may then wait a long, random time between plays, and the length of that wait depends on the arm's posterior, which in turn depends on how long it has waited. Counting plays of the suboptimal arm with a union bound over rounds, under the assumption that the optimal arm is already concentrated, therefore does not work; controlling these waiting times is the central difficulty and is where the 1/Δ31/\Delta^31/Δ3 dependence enters.

Formalization scope

  • Model. The instance is the platform's StochasticBandit 2 (a probability measure on R\mathbb RR per arm, mean banditArmMean), with the hypothesis that each reward law gives mass 111 to [0,1][0,1][0,1]. Lean arm 0 is the paper's arm 1 and Lean arm 1 the paper's arm 2. Lean rounds are indexed from 000.
  • Algorithm. Algorithm 2 is realized on one probability space with three independent i.i.d. tables: Beta draws W(i,t,a,b)∼Beta(a+1,b+1)W(i,t,a,b)\sim\mathrm{Beta}(a+1,b+1)W(i,t,a,b)∼Beta(a+1,b+1), rewards X(i,t)∼DiX(i,t)\sim D_iX(i,t)∼Di​, and uniforms V(i,t)V(i,t)V(i,t). Round ttt uses θi(t)=W(i,t,Si(t),Fi(t))\theta_i(t)=W(i,t,S_i(t),F_i(t))θi​(t)=W(i,t,Si​(t),Fi​(t)), r~t=X(i(t),t)\tilde r_t=X(i(t),t)r~t​=X(i(t),t) and rt=1{V(i(t),t)<r~t}r_t=\mathbf 1\{V(i(t),t)<\tilde r_t\}rt​=1{V(i(t),t)<r~t​}. Ties in the arg max go to the smaller index (a null event).
  • Values. Regret and expectations are lower Lebesgue integrals in [0,∞][0,\infty][0,∞]. X(j,s,y)X(j,s,y)X(j,s,y) is N∪{∞}\mathbb N\cup\{\infty\}N∪{∞}-valued, so Lemma 1 at y=1y=1y=1 reads ∞=∞\infty=\infty∞=∞, as in the paper.
  • O(·). The paper's O(⋅)O(\cdot)O(⋅) (footnote 1: f≤cgf\le cgf≤cg for n≥n0n\ge n_0n≥n0​) is stated with one universal constant C>0C>0C>0, quantified before the instance, the means and the horizon, for all T≥2T\ge 2T≥2. Eq. (1) is stated the same way, without its printed numerals.
  • Not trivial. The goal is about Algorithm 2 itself, with fresh Beta samples, fresh rewards and the Bernoulli coin. A statement about "any policy satisfying Lemma 2's event bound", or one whose constant depends on Δ\DeltaΔ, the reward laws or TTT, would not be Theorem 1.
  • Edge cases. μ1<1\mu_1<1μ1​<1 is assumed only in Lemma 3, where the paper's RRR and DDD require it. It is not a hypothesis of the goal.
  • Infrastructure. A complete proof needs: inverse-transform or order-statistics facts for Beta laws (Fact 1); geometric expectations; Hoeffding's inequality for sums of Bernoulli variables (Mathlib has Hoeffding/Azuma); the binomial median theorem (Jogdeo–Samuels; Kaas–Buhrman); and the coupling from the reward tables to the per-arm i.i.d. output stacks the paper reasons with. Fact 1, Lemma 6 and Fact 2 are reusable beyond this mission. Contributions to any milestone are welcome, and so is a direct proof of the regret bound with an explicit constant.

Selected references

  • S. Agrawal and N. Goyal, Analysis of Thompson Sampling for the Multi-armed Bandit Problem, COLT 2012; arXiv:1111.1797v3. https://arxiv.org/abs/1111.1797
  • W. R. Thompson, On the likelihood that one unknown probability exceeds another in view of the evidence of two samples, Biometrika 25 (1933) 285–294. https://doi.org/10.2307/2332286
  • T. L. Lai and H. Robbins, Asymptotically efficient adaptive allocation rules, Advances in Applied Mathematics 6 (1985) 4–22. https://doi.org/10.1016/0196-8858(85)90002-8
  • P. Auer, N. Cesa-Bianchi and P. Fischer, Finite-time analysis of the multiarmed bandit problem, Machine Learning 47 (2002) 235–256. https://doi.org/10.1023/A:1013689704352
  • K. Jogdeo and S. M. Samuels, Monotone convergence of binomial probabilities and a generalization of Ramanujan's equation, Annals of Mathematical Statistics 39 (1968) 1191–1195. https://doi.org/10.1214/aoms/1177698243
  • R. Kaas and J. M. Buhrman, Mean, median and mode in binomial distributions, Statistica Neerlandica 34 (1980) 13–18. https://doi.org/10.1111/j.1467-9574.1980.tb00681.x
  • O. Chapelle and L. Li, An empirical evaluation of Thompson Sampling, NIPS 2011. https://papers.nips.cc/paper/4321-an-empirical-evaluation-of-thompson-sampling
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Optimal Best Arm Identification with Fixed Confidence III: δ-PAC Guarantee of Chernoff's Stopping Rule for Bernoulli BanditsResearch Paper

Motivation

In best arm identification with fixed confidence, a learner samples KKK unknown distributions ("arms") one at a time, and must eventually stop and name the arm with the largest mean, with an error probability at most a prescribed risk δ\deltaδ, while using as few samples as possible. The problem goes back to the sequential design of experiments (Chernoff, 1959; Even-Dar, Mannor and Mansour, 2006) and underlies adaptive A/B testing, clinical trial design and hyperparameter selection.

Any fixed-confidence strategy consists of three parts: a sampling rule, a stopping rule and a decision rule. Garivier and Kaufmann (arXiv:1602.04589, COLT 2016) proposed the Track-and-Stop strategy, the first shown to match the asymptotic lower bound on the expected sample complexity. Its stopping rule is a generalized likelihood ratio (GLR) test, Chernoff's stopping rule. Its correctness, the guarantee that the recommended arm is wrong with probability at most δ\deltaδ, must hold whatever the sampling rule, which is what allows the sampling rule to be tuned freely for efficiency. This mission formalizes that guarantee for Bernoulli arms, Theorem 10 of the paper, with its explicit threshold β(t,δ)=log⁡(2t(K−1)/δ)\beta(t,\delta) = \log(2t(K-1)/\delta)β(t,δ)=log(2t(K−1)/δ).

Setting

The arms are A={1,…,K}\mathcal A = \{1,\dots,K\}A={1,…,K}. A Bernoulli bandit model is a mean vector μ=(μ1,…,μK)∈[0,1]K\boldsymbol\mu = (\mu_1,\dots,\mu_K) \in [0,1]^Kμ=(μ1​,…,μK​)∈[0,1]K: pulling arm aaa returns reward 111 with probability μa\mu_aμa​ and 000 otherwise, independently of the past. The class S\mathcal SS contains the models with a unique optimal arm a∗(μ)a^*(\boldsymbol\mu)a∗(μ), i.e. μa∗>μi\mu_{a^*} > \mu_iμa∗​>μi​ for all i≠a∗i \ne a^*i=a∗.

At round ttt the learner chooses an arm AtA_tAt​ as a (possibly randomized) function of the past observations and observes a reward XtX_tXt​. Write Na(t)N_a(t)Na​(t) for the number of pulls of arm aaa among the first ttt rounds, sa(t)s_a(t)sa​(t) for the number of those pulls that returned 111, and μ^a(t)=Na(t)−1∑s≤tXs1{As=a}\hat\mu_a(t) = N_a(t)^{-1}\sum_{s \le t} X_s \mathbb 1\{A_s = a\}μ^​a​(t)=Na​(t)−1∑s≤t​Xs​1{As​=a} for the empirical mean. The likelihood of arm aaa's observations under mean uuu is pu(X‾Na(t)a)=usa(t)(1−u)Na(t)−sa(t)p_u(\underline X^a_{N_a(t)}) = u^{s_a(t)}(1-u)^{N_a(t)-s_a(t)}pu​(X​Na​(t)a​)=usa​(t)(1−u)Na​(t)−sa​(t).

The GLR statistic for "arm aaa is at least as good as arm bbb" is

Za,b(t)=log⁡max⁡μa′≥μb′pμa′(X‾Na(t)a) pμb′(X‾Nb(t)b)max⁡μa′≤μb′pμa′(X‾Na(t)a) pμb′(X‾Nb(t)b).Z_{a,b}(t) = \log \frac{\max_{\mu'_a \ge \mu'_b} p_{\mu'_a}(\underline X^a_{N_a(t)})\, p_{\mu'_b}(\underline X^b_{N_b(t)})}{\max_{\mu'_a \le \mu'_b} p_{\mu'_a}(\underline X^a_{N_a(t)})\, p_{\mu'_b}(\underline X^b_{N_b(t)})}.Za,b​(t)=logmaxμa′​≤μb′​​pμa′​​(X​Na​(t)a​)pμb′​​(X​Nb​(t)b​)maxμa′​≥μb′​​pμa′​​(X​Na​(t)a​)pμb′​​(X​Nb​(t)b​)​.

Chernoff's stopping rule with exploration rate β(t,δ)\beta(t,\delta)β(t,δ) is

τδ=inf⁡{t≥1:∃a∈A, ∀b≠a, Za,b(t)>β(t,δ)},\tau_\delta = \inf\{t \ge 1 : \exists a \in \mathcal A,\ \forall b \ne a,\ Z_{a,b}(t) > \beta(t,\delta)\},τδ​=inf{t≥1:∃a∈A, ∀b=a, Za,b​(t)>β(t,δ)},

and the decision rule recommends a^τδ∈argmax⁡aμ^a(τδ)\hat a_{\tau_\delta} \in \operatorname{argmax}_a \hat\mu_a(\tau_\delta)a^τδ​​∈argmaxa​μ^​a​(τδ​).

The Krichevsky–Trofimov (KT) distribution on binary sequences x∈{0,1}nx \in \{0,1\}^nx∈{0,1}n is kt(x)=∫01(πu(1−u))−1pu(x) du\mathrm{kt}(x) = \int_0^1 \big(\pi\sqrt{u(1-u)}\big)^{-1} p_u(x)\,\mathrm dukt(x)=∫01​(πu(1−u)​)−1pu​(x)du, the Bernoulli likelihood mixed over the Beta(1/2,1/2)(1/2,1/2)(1/2,1/2) prior.

Formalization targets

Goal: Theorem 10

For every δ∈(0,1)\delta \in (0,1)δ∈(0,1), every sampling strategy, and the threshold β(t,δ)=log⁡(2t(K−1)/δ)\beta(t,\delta) = \log\big(2t(K-1)/\delta\big)β(t,δ)=log(2t(K−1)/δ),

∀μ∈S,Pμ(τδ<∞, a^τδ≠a∗)≤δ.\forall \boldsymbol\mu \in \mathcal S,\qquad \mathbb P_{\boldsymbol\mu}\big(\tau_\delta < \infty,\ \hat a_{\tau_\delta} \ne a^*\big) \le \delta .∀μ∈S,Pμ​(τδ​<∞, a^τδ​​=a∗)≤δ.

Milestone: Lemma 11 (Willems, Shtarkov and Tjalkens, 1995)

kt\mathrm{kt}kt is a probability law on {0,1}n\{0,1\}^n{0,1}n, and for n≥1n \ge 1n≥1,

sup⁡x∈{0,1}n sup⁡u∈[0,1]pu(x)kt(x)≤2n.\sup_{x\in\{0,1\}^n}\ \sup_{u\in[0,1]} \frac{p_u(x)}{\mathrm{kt}(x)} \le 2\sqrt n .x∈{0,1}nsup​ u∈[0,1]sup​kt(x)pu​(x)​≤2n​.

Milestone: the pairwise crossing bound of Appendix C.1

With Ta,b=inf⁡{t:Za,b(t)>β(t,δ)}T_{a,b} = \inf\{t : Z_{a,b}(t) > \beta(t,\delta)\}Ta,b​=inf{t:Za,b​(t)>β(t,δ)}, for all arms with μa<μb\mu_a < \mu_bμa​<μb​,

Pμ(Ta,b<∞)≤δK−1.\mathbb P_{\boldsymbol\mu}(T_{a,b} < \infty) \le \frac{\delta}{K-1}.Pμ​(Ta,b​<∞)≤K−1δ​.

Significance

Theorem 10 decouples correctness from efficiency. Because the guarantee holds for every sampling strategy, any sampling rule, including the C-Tracking and D-Tracking rules of Track-and-Stop, the uniform rule, or a heuristic, inherits δ\deltaδ-correctness as soon as it is paired with Chernoff's stopping rule at this threshold. The asymptotic optimality result of the paper (Theorem 14) then only has to control the sample complexity. The threshold is explicit, with no unspecified constant, in contrast to the deviational threshold of Proposition 12.

The result is proved in the paper, in Appendix C.1, and rests on Lemma 11, which the paper quotes from the universal coding literature without proof. As far as the platform's catalog shows, none of these results is formalized. The platform holds a machine-checkable statement of the analogous result for Gaussian arms with the Lattimore–Szepesvári threshold (BanditAlgorithm.chernoff_stopping_rule_sound, Lemma 33.7 of Bandit Algorithms), which is a different model and a different threshold. Formalizing Theorem 10 adds a proof of Lemma 11 (the KT regret bound, reusable in information theory and universal prediction), the Bernoulli GLR statistic, and a change of measure from the true bandit law to a Bayesian mixture law on the trajectory space.

Difficulty

The obvious approach bounds, for each fixed ttt, the probability that Za,b(t)Z_{a,b}(t)Za,b​(t) exceeds β(t,δ)\beta(t,\delta)β(t,δ) by a concentration inequality and sums over ttt. This fails: the sampling strategy is arbitrary and adaptive, so Na(t)N_a(t)Na​(t) and Nb(t)N_b(t)Nb​(t) are random and depend on the past rewards, and a fixed-sample-size deviation bound does not apply; a union bound over the possible values of the counts loses more than the threshold allows. The maximum likelihood in the numerator of Za,bZ_{a,b}Za,b​ is also not a probability density, so the likelihood ratio cannot directly be read as a change of measure. The argument must control the whole trajectory law under an arbitrary randomized policy, and must handle empty samples (an arm never pulled contributes likelihood 111) and the boundary means 000 and 111.

Formalization scope

The formalization is in Lean 4 with Mathlib and reuses the platform's canonical bandit model: StochasticBandit, BanditPolicy (a Markov kernel per round from the observed history to the next arm, so randomized strategies are included), banditTrajMeasure (the law of the infinite trajectory, where coordinate sss is round s+1s+1s+1) and IsSoundBAI from BanditTrajectory; bernoulliBandit from bernoulliRelativeEntropy; and only the pull counts trajPullCount and empirical means trajEmpiricalMean from TrackAndStop. The Gaussian GLR and threshold of TrackAndStop are not used.

Conventions committed to:

  • Arms are Fin K. Bernoulli means range over [0,1][0,1][0,1], degenerate laws included; the paper's exponential-family mean space is (0,1)(0,1)(0,1), so the [0,1][0,1][0,1] statement implies the paper's.
  • Za,b(t)Z_{a,b}(t)Za,b​(t) is defined as the ratio of the two maxima over [0,1]2[0,1]^2[0,1]2, not by the closed form (7), which holds only when μ^a(t)≥μ^b(t)\hat\mu_a(t) \ge \hat\mu_b(t)μ^​a​(t)≥μ^​b​(t). Both maxima are attained and positive.
  • The stopping rule ranges over t≥1t \ge 1t≥1; at t=0t = 0t=0 there is no observation and the paper's β(0,δ)=log⁡0\beta(0,\delta) = \log 0β(0,δ)=log0 is undefined. τδ=∞\tau_\delta = \inftyτδ​=∞ when the rule never fires.
  • The decision rule is quantified: the goal holds for every recommendation that maximizes the empirical mean at τδ\tau_\deltaτδ​, whatever the tie-breaking.
  • Probabilities of events are outer measures under the trajectory law; no measurability is assumed.
  • K≥1K \ge 1K≥1 only. For K=1K = 1K=1 the statement is trivially true (no suboptimal arm).

Disclosed deviations from the page: Lemma 11's ratio bound is stated for n≥1n \ge 1n≥1, since at n=0n = 0n=0 the printed bound reads 1≤01 \le 01≤0; the use of the lemma in Appendix C.1 is unaffected. Appendix C.1 calls the result "Proposition 10" (a slip for Theorem 10) and prints the KT density as 1/πu(1−u)1/\sqrt{\pi u(1-u)}1/πu(1−u)​ (a slip for 1/(πu(1−u))1/(\pi\sqrt{u(1-u)})1/(πu(1−u)​) of Lemma 11, which is the normalized one); the formalization follows Lemma 11.

A trivializing formalization is ruled out: stating the result for Gaussian arms or with the Lattimore–Szepesvári threshold is the platform's existing Lemma 33.7 and is not Theorem 10, and a free decision rule without the argmax hypothesis would make the claim false rather than faithful.

Contributions welcome: a proof of Lemma 11; a change-of-measure lemma for banditTrajMeasure under a mixture of environments; and the union-bound reduction from the goal to the pairwise claim.

Selected references

  • A. Garivier, E. Kaufmann, Optimal Best Arm Identification with Fixed Confidence, COLT 2016 (JMLR W&CP 49), arXiv:1602.04589v2, 2016. https://arxiv.org/abs/1602.04589
  • F. M. J. Willems, Y. M. Shtarkov, T. J. Tjalkens, The context-tree weighting method: basic properties, IEEE Transactions on Information Theory 41(3), 1995. https://doi.org/10.1109/18.382012
  • R. Krichevsky, V. Trofimov, The performance of universal encoding, IEEE Transactions on Information Theory 27(2), 1981. https://doi.org/10.1109/TIT.1981.1056331
  • H. Chernoff, Sequential design of experiments, Annals of Mathematical Statistics 30(3), 1959. https://doi.org/10.1214/aoms/1177706205
  • T. Lattimore, C. Szepesvári, Bandit Algorithms, Cambridge University Press, 2020, Chapter 33. https://doi.org/10.1017/9781108571401
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On the Uniform Convergence of Relative Frequencies of Events to Their Probabilities III: Uniform Convergence and the Entropy per ObservationResearch Paper

Motivation

Estimating a probability by the relative frequency of the event in an independent sample is justified for one event by the law of large numbers. Statistics and learning theory need more: the frequencies of a whole class of events SSS must approach their probabilities simultaneously, so that a quantity chosen after looking at the data (the empirical risk minimizer, the empirical distribution function) is still close to its expectation. Glivenko's theorem on the empirical distribution function is the classical instance; empirical risk minimization rests on the same property for the class of loss sets of a model.

Vapnik and Chervonenkis, On the Uniform Convergence of Relative Frequencies of Events to Their Probabilities, Theory Probab. Appl. 16 (1971), treat this question in two parts. The first gives a distribution-free sufficient condition through the growth function (Theorems 1–3). The second, which this mission formalizes, gives a condition that is necessary and sufficient for a fixed distribution: Theorem 4, the entropy criterion.

Timeline. 1933: Glivenko and Cantelli prove uniform convergence for the class of rays {x≤a}\{x \le a\}{x≤a} on the line. 1968: Vapnik and Chervonenkis announce the results in Dokl. Akad. Nauk SSSR 181. 1971: the full paper appears, with the growth-function bound and the entropy criterion. Later work (Talagrand 1987; Dudley, Giné and Zinn 1991) recasts such criteria as the theory of Glivenko–Cantelli classes.

Setting

Let XXX be a set carrying a probability measure PPP, and SSS a collection of measurable subsets of XXX (events). A sample of size lll is a sequence x1,…,xlx_1, \dots, x_lx1​,…,xl​ of independent draws from PPP; repetitions are allowed. For A∈SA \in SA∈S the relative frequency νA(l)\nu_A^{(l)}νA(l)​ is the fraction of sample terms lying in AAA, and PA=P(A)P_A = P(A)PA​=P(A). The maximal deviation is

π(l)(x1,…,xl)=sup⁡A∈S∣νA(l)−PA∣.\pi^{(l)}(x_1, \dots, x_l) = \sup_{A \in S} \bigl|\nu_A^{(l)} - P_A\bigr| .π(l)(x1​,…,xl​)=A∈Ssup​​νA(l)​−PA​​.

The relative frequencies converge in probability to the probabilities uniformly over SSS when P{π(l)>ε}→0\mathbf{P}\{\pi^{(l)} > \varepsilon\} \to 0P{π(l)>ε}→0 as l→∞l \to \inftyl→∞ for every ε>0\varepsilon > 0ε>0.

Each A∈SA \in SA∈S induces in a sample the subsample of terms lying in AAA. The index ΔS(x1,…,xl)\Delta^S(x_1, \dots, x_l)ΔS(x1​,…,xl​) is the number of different subsamples induced by the sets of SSS; it lies between 000 and 2l2^l2l. The entropy of SSS in samples of size lll is

HS(l)=Elog⁡2ΔS(x1,…,xl).H^S(l) = \mathbf{E} \log_2 \Delta^S(x_1, \dots, x_l) .HS(l)=Elog2​ΔS(x1​,…,xl​).

For a sample of size 2l2l2l, split into halves x1,…,xlx_1, \dots, x_lx1​,…,xl​ and xl+1,…,x2lx_{l+1}, \dots, x_{2l}xl+1​,…,x2l​ with relative frequencies νA′\nu'_AνA′​ and νA′′\nu''_AνA′′​, the semi-sample deviation is ρ(l)=sup⁡A∈S∣νA′−νA′′∣\rho^{(l)} = \sup_{A \in S} |\nu'_A - \nu''_A|ρ(l)=supA∈S​∣νA′​−νA′′​∣. Finally Φ(n,r)\Phi(n, r)Φ(n,r) is defined by the recurrence Φ(n,r)=Φ(n,r−1)+Φ(n−1,r−1)\Phi(n, r) = \Phi(n, r-1) + \Phi(n-1, r-1)Φ(n,r)=Φ(n,r−1)+Φ(n−1,r−1), Φ(0,r)=Φ(n,0)=1\Phi(0, r) = \Phi(n, 0) = 1Φ(0,r)=Φ(n,0)=1.

Formalization targets

Goal: Theorem 4 (p. 275)

(∀ε>0: lim⁡l→∞P{π(l)>ε}=0)  ⟺  lim⁡l→∞HS(l)l=0.\Bigl(\forall \varepsilon > 0:\ \lim_{l\to\infty} \mathbf{P}\{\pi^{(l)} > \varepsilon\} = 0\Bigr) \iff \lim_{l \to \infty} \frac{H^S(l)}{l} = 0 .(∀ε>0: l→∞lim​P{π(l)>ε}=0)⟺l→∞lim​lHS(l)​=0.

Milestones

  1. Entropy rate. (12) ΔS(x1,…,xl)≤ΔS(x1,…,xk)ΔS(xk+1,…,xl)\Delta^S(x_1, \dots, x_l) \le \Delta^S(x_1, \dots, x_k)\Delta^S(x_{k+1}, \dots, x_l)ΔS(x1​,…,xl​)≤ΔS(x1​,…,xk​)ΔS(xk+1​,…,xl​); the subadditivity HS(l1+l2)≤HS(l1)+HS(l2)H^S(l_1 + l_2) \le H^S(l_1) + H^S(l_2)HS(l1​+l2​)≤HS(l1​)+HS(l2​); Lemma 3, HS(l)/l→c∈[0,1]H^S(l)/l \to c \in [0, 1]HS(l)/l→c∈[0,1]; Lemma 4, P(∣l−1log⁡2ΔS−c∣>ε)→0\mathbf{P}(|l^{-1}\log_2 \Delta^S - c| > \varepsilon) \to 0P(∣l−1log2​ΔS−c∣>ε)→0.
  2. Sufficiency. Lemma 2, P{π(l)>ε}≤2 P{ρ(l)≥ε/2}\mathbf{P}\{\pi^{(l)} > \varepsilon\} \le 2\,\mathbf{P}\{\rho^{(l)} \ge \varepsilon/2\}P{π(l)>ε}≤2P{ρ(l)≥ε/2} for l≥2/ε2l \ge 2/\varepsilon^2l≥2/ε2; the per-sample permutation bound 2ΔS(x1,…,x2l)e−ε2l/82\Delta^S(x_1, \dots, x_{2l}) e^{-\varepsilon^2 l/8}2ΔS(x1​,…,x2l​)e−ε2l/8; and
P{ρ(l)≥ε2}≤2(2e)ε2l/8+P{12llog⁡2ΔS(x1,…,x2l)>ε216}.\mathbf{P}\{\rho^{(l)} \ge \tfrac{\varepsilon}{2}\} \le 2\Bigl(\frac{2}{e}\Bigr)^{\varepsilon^2 l/8} + \mathbf{P}\Bigl\{\tfrac{1}{2l}\log_2 \Delta^S(x_1, \dots, x_{2l}) > \tfrac{\varepsilon^2}{16}\Bigr\} .P{ρ(l)≥2ε​}≤2(e2​)ε2l/8+P{2l1​log2​ΔS(x1​,…,x2l​)>16ε2​}.
  1. Necessity. Lemma 1 (Sauer–Shelah in sequence form); step 1°, 1−P(C′)≥(1−P(Q))21 - \mathbf{P}(C') \ge (1 - \mathbf{P}(Q))^21−P(C′)≥(1−P(Q))2 with C′={ρ(l)>2ε}C' = \{\rho^{(l)} > 2\varepsilon\}C′={ρ(l)>2ε}; (26), P{ΔS>Φ([ql],l)}→1\mathbf{P}\{\Delta^S > \Phi([ql], l)\} \to 1P{ΔS>Φ([ql],l)}→1 when 0<q<140 < q < \frac140<q<41​ and qlog⁡2(2e/q)<cq\log_2(2e/q) < cqlog2​(2e/q)<c; and (29), P{π(l)>ε}→1\mathbf{P}\{\pi^{(l)} > \varepsilon\} \to 1P{π(l)>ε}→1 when moreover 0<ε<q/70 < \varepsilon < q/70<ε<q/7.

Significance

Theorem 4 characterizes uniform convergence for a given distribution exactly, with no gap between the necessary and the sufficient condition. It separates the cases the growth-function bound cannot: a class may have mS(l)=2lm^S(l) = 2^lmS(l)=2l for every lll (all open subsets of [0,1][0,1][0,1]) and still satisfy HS(l)/l→0H^S(l)/l \to 0HS(l)/l→0 under a particular PPP, or fail it. The entropy HS(l)H^S(l)HS(l) is the distribution-dependent quantity from which later work on Glivenko–Cantelli classes and on consistency of empirical risk minimization proceeds; the 1981 paper of the same authors extends the criterion to classes of functions. The quantitative form (29) states more than the negation of convergence: when the entropy rate is positive, the maximal deviation stays above a fixed ε\varepsilonε with probability tending to one.

The result has been proved since 1971; it has not been formalized. The platform holds Sauer–Shelah variants over sets of distinct points and PAC bounds with other constants, but no statement of the VC entropy or of Theorem 4. The mission produces machine-checked statements of the entropy criterion and of its supporting lemmas with the paper's own constants (l≥2/ε2l \ge 2/\varepsilon^2l≥2/ε2, 2e−ε2l/82e^{-\varepsilon^2 l/8}2e−ε2l/8, δ=ε2/16\delta = \varepsilon^2/16δ=ε2/16, ε<q/7\varepsilon < q/7ε<q/7).

Difficulty

The sufficiency half is a variant of the proof of the growth-function bound; its new ingredient is the concentration of l−1log⁡2ΔSl^{-1} \log_2 \Delta^Sl−1log2​ΔS (Lemma 4), which needs subadditivity and a law of large numbers over independent blocks of the sample rather than a single mean estimate. The hypergeometric tail estimate behind the permutation bound is omitted in the paper ("a simple but long computation").

Necessity is harder. The obvious attempt, bounding P{π(l)>ε}\mathbf{P}\{\pi^{(l)} > \varepsilon\}P{π(l)>ε} from below by exhibiting a single bad event, fails: SSS may be uncountable and no single AAA deviates with non-vanishing probability. A positive entropy rate has to be converted into a combinatorial statement about typical samples ((26) combines Lemma 4 with an estimate of Φ([ql],l)\Phi([ql], l)Φ([ql],l)), and that statement back into a lower bound on a probability over the product measure; the constants q<14q < \frac14q<41​ and ε<q/7\varepsilon < q/7ε<q/7 must be tracked through both conversions, and the conclusion lim⁡P{π(l)>ε}=1\lim \mathbf{P}\{\pi^{(l)} > \varepsilon\} = 1limP{π(l)>ε}=1 needs the unweakened inequality of step 1°.

Formalization scope

A sample of size lll is a function Fin l → X (positions 0,…,l−10, \dots, l-10,…,l−1) and its law is the product measure Measure.pi (fun _ => P), with P a probability measure. A subsample is a set of positions, so the index counts distinct Finset (Fin l) of the form {i:xi∈A}\{i : x_i \in A\}{i:xi​∈A}. The halves of x : Fin (l + l) → X are x ∘ Fin.castAdd l and x ∘ Fin.natAdd l. PAP_APA​ is P.real A; the suprema π(l)\pi^{(l)}π(l) and ρ(l)\rho^{(l)}ρ(l) are real suprema over the subtype of SSS (values in [0,1][0,1][0,1]; 000 for S=∅S = \emptysetS=∅). HS(l)H^S(l)HS(l) is a Bochner integral of Real.logb 2 of the index, and [ql][ql][ql] is ⌊q * l⌋₊. Probabilities are values in [0,∞][0, \infty][0,∞], except in the inequalities between probabilities (step 1°, the sufficiency estimate), which use Measure.real.

Measurability. The paper assumes, and the statements carry as hypotheses, that the events of SSS are measurable (p. 264), that π(l)\pi^{(l)}π(l) is a random variable (p. 265), that ρ(l)\rho^{(l)}ρ(l) is measurable (p. 268), and that the index is measurable in the sample (p. 273). Each statement carries the ones its proof uses. Without them the Bochner integral defining HS(l)H^S(l)HS(l) can be the junk value 000 and the equivalence can fail; replacing them by "SSS countable" would weaken the theorem. The goal is not trivialized by degenerate cases: the equivalence is not vacuous for any class, and S=∅S = \emptysetS=∅ gives the true instance HS=0H^S = 0HS=0, π(l)=0\pi^{(l)} = 0π(l)=0.

Corrections of the printed text. Lemma 2 is printed for l>2/ε2l > 2/\varepsilon^2l>2/ε2; its proof gives l≥2/ε2l \ge 2/\varepsilon^2l≥2/ε2, which is stated. On p. 275 Lemma 2 is recalled as "2P(C)≥12P(Q)2\mathbf{P}(C) \ge \frac12 P(Q)2P(C)≥21​P(Q)", meaning P(C)≥12P(Q)\mathbf{P}(C) \ge \frac12\mathbf{P}(Q)P(C)≥21​P(Q). On p. 276 the first display carries a stray upper limit "4" on the integral, and the region of integration is printed {log⁡2ΔS≤2δ}\{\log_2 \Delta^S \le 2\delta\}{log2​ΔS≤2δ} where {log⁡2ΔS≤2δl}\{\log_2\Delta^S \le 2\delta l\}{log2​ΔS≤2δl} is meant. The event C′C'C′ is defined with ">2ε> 2\varepsilon>2ε" (p. 276) but integrated in step 3° as θ(⋅−2ε)\theta(\cdot - 2\varepsilon)θ(⋅−2ε), which counts "≥2ε\ge 2\varepsilon≥2ε"; the strict form is stated, and the estimate of step 3° is itself strict. Step 1° is stated unweakened. Milestone texts are verbatim.

Contributions welcome: a reusable development of the index and its submultiplicativity, the hypergeometric tail bound for sampling without replacement, a block law of large numbers for subadditive functionals of i.i.d. samples, and the permutation-invariance argument for product measures on Fin (l + l) → X.

Selected references

  • V. N. Vapnik and A. Ya. Chervonenkis, On the uniform convergence of relative frequencies of events to their probabilities, Theory of Probability and Its Applications 16(2) (1971), 264–280. https://doi.org/10.1137/1116025
  • V. N. Vapnik and A. Ya. Chervonenkis, Necessary and sufficient conditions for the uniform convergence of means to their expectations, Theory of Probability and Its Applications 26(3) (1981), 532–553. https://doi.org/10.1137/1126059
  • M. Talagrand, The Glivenko–Cantelli problem, Annals of Probability 15(3) (1987), 837–870. https://doi.org/10.1214/aop/1176992069
  • R. M. Dudley, E. Giné and J. Zinn, Uniform and universal Glivenko–Cantelli classes, Journal of Theoretical Probability 4(3) (1991), 485–510. https://doi.org/10.1007/BF01210321
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Operations ResearchOptimizationProbability+1·Captain: mikedeng1

A Distributional Interpretation of Robust Optimization II: Box-Robust Sample Average Optimization Is ConsistentResearch Paper

Why robustify a sampled stochastic program

Many decision problems under uncertainty take the form of a stochastic program: choose a decision vvv from a feasible set F\mathcal FF to maximise the expected utility Ex∼μ[f(v,x)]\mathbb E_{x\sim\mu}[f(v,x)]Ex∼μ​[f(v,x)], where the distribution μ\muμ of the uncertain parameter x∈Rmx\in\mathbb R^mx∈Rm is known only through i.i.d. samples x1,…,xnx_1,\dots,x_nx1​,…,xn​. The standard remedy, sample average approximation, maximises 1n∑if(v,xi)\frac1n\sum_i f(v,x_i)n1​∑i​f(v,xi​) instead. Its consistency (convergence of the optimal expected utility of its solutions to the true optimum) is classical, but it needs regularity assumptions of its own, for example those of King and Wets (Stochastics and Stochastic Reports, 1991), cited on p. 98 of the paper; the paper presents its construction as a route to consistency under weaker conditions.

Robust optimization (RO) takes a different route: it protects each sample by an uncertainty set and optimises against the worst point in it. Xu, Caramanis and Mannor (Math. Oper. Res. 2012) show that RO over several overlapping uncertainty sets is equivalent to a distributionally robust stochastic program (their Theorem 2.1, the subject of mission I of this series). Section 3 of the paper uses that equivalence to show that a specific robustification of the sampled problem, with ℓ∞\ell_\inftyℓ∞​ boxes of shrinking radius around each sample, is consistent under only boundedness and equicontinuity of fff. This mission formalizes that result, Theorem 3.1.

Setting

Equip Rm\mathbb R^mRm with the sup norm ∥z∥∞=max⁡k∣zk∣\|z\|_\infty=\max_k|z_k|∥z∥∞​=maxk​∣zk​∣, its Borel σ\sigmaσ-algebra and Lebesgue measure dxdxdx. The data are:

  • a set of decisions VVV and a nonempty feasible set F⊆V\mathcal F\subseteq VF⊆V;
  • a utility f:V×Rm→Rf:V\times\mathbb R^m\to\mathbb Rf:V×Rm→R, Borel measurable in xxx for each vvv;
  • a true density h∗h^*h∗ on Rm\mathbb R^mRm (nonnegative, ∫h∗ dx=1\int h^*\,dx=1∫h∗dx=1) and i.i.d. samples x1,x2,…x_1,x_2,\dotsx1​,x2​,… with distribution h∗(x) dxh^*(x)\,dxh∗(x)dx;
  • radii ϵ(n)>0\epsilon(n)>0ϵ(n)>0.

For a sample x1,…,xnx_1,\dots,x_nx1​,…,xn​ the boxes are Zi={xi+δ∣∥δ∥∞≤ϵ(n)}\mathcal Z_i=\{x_i+\delta\mid\|\delta\|_\infty\le\epsilon(n)\}Zi​={xi​+δ∣∥δ∥∞​≤ϵ(n)}, and the box-robust sample objective is

Jn(v)=1n∑i=1n inf⁡∥δi∥∞≤ϵ(n)f(v,xi+δi)=∑i=1n1ninf⁡xi′∈Zif(v,xi′).J_n(v)=\frac1n\sum_{i=1}^n\ \inf_{\|\delta_i\|_\infty\le\epsilon(n)}f(v,x_i+\delta_i)=\sum_{i=1}^n\frac1n\inf_{x_i'\in\mathcal Z_i}f(v,x_i').Jn​(v)=n1​i=1∑n​ ∥δi​∥∞​≤ϵ(n)inf​f(v,xi​+δi​)=i=1∑n​n1​xi′​∈Zi​inf​f(v,xi′​).

The RO solution v(n)v(n)v(n) is a maximiser of JnJ_nJn​ over F\mathcal FF. The equicontinuity modulus of fff is

d(ϵ)=sup⁡v, x, ∥δ∥∞≤ϵ∣f(v,x)−f(v,x+δ)∣.d(\epsilon)=\sup_{v,\,x,\ \|\delta\|_\infty\le\epsilon}|f(v,x)-f(v,x+\delta)|.d(ϵ)=v,x, ∥δ∥∞​≤ϵsup​∣f(v,x)−f(v,x+δ)∣.

The proof works with the distribution set Pn\mathcal P_nPn​ of probability measures μ\muμ with μ(⋃i∈SZi)≥∣S∣/n\mu(\bigcup_{i\in S}\mathcal Z_i)\ge|S|/nμ(⋃i∈S​Zi​)≥∣S∣/n for every S⊆{1,…,n}S\subseteq\{1,\dots,n\}S⊆{1,…,n}, and with the uniform box kernel density estimator

hn(x)=(nϵ(n)m)−1∑i=1nK(x−xiϵ(n)),K(z)=1(∥z∥∞≤1)2m.h_n(x)=(n\epsilon(n)^m)^{-1}\sum_{i=1}^nK\Big(\frac{x-x_i}{\epsilon(n)}\Big),\qquad K(z)=\frac{\mathbf 1(\|z\|_\infty\le1)}{2^m}.hn​(x)=(nϵ(n)m)−1i=1∑n​K(ϵ(n)x−xi​​),K(z)=2m1(∥z∥∞​≤1)​.

Formalization targets

Goal: Theorem 3.1 (p. 98)

Assume ∣f(v,x)∣≤C|f(v,x)|\le C∣f(v,x)∣≤C for all v,xv,xv,x; d(ϵ)→0d(\epsilon)\to0d(ϵ)→0 as ϵ↓0\epsilon\downarrow0ϵ↓0; ϵ(n)↓0\epsilon(n)\downarrow0ϵ(n)↓0 and nϵ(n)m↑∞n\epsilon(n)^m\uparrow\inftynϵ(n)m↑∞. Then for every choice of maximisers v(n)v(n)v(n), with probability one,

lim⁡n→∞∫Rmf(v(n),x) h∗(x) dx=sup⁡v∈F∫Rmf(v,x) h∗(x) dx.\lim_{n\to\infty}\int_{\mathbb R^m}f(v(n),x)\,h^*(x)\,dx=\sup_{v\in\mathcal F}\int_{\mathbb R^m}f(v,x)\,h^*(x)\,dx .n→∞lim​∫Rm​f(v(n),x)h∗(x)dx=v∈Fsup​∫Rm​f(v,x)h∗(x)dx.

Milestones (proof of Theorem 3.1, p. 99)

  1. hnh_nhn​ is the density of a probability measure in Pn\mathcal P_nPn​.
  2. Jn(v)≤∫f(v,x) hn(x) dxJ_n(v)\le\int f(v,x)\,h_n(x)\,dxJn​(v)≤∫f(v,x)hn​(x)dx for every vvv.
  3. Oscillation over a box: sup⁡Zif(v,⋅)−inf⁡Zif(v,⋅)≤d(2ϵ(n))\sup_{\mathcal Z_i}f(v,\cdot)-\inf_{\mathcal Z_i}f(v,\cdot)\le d(2\epsilon(n))supZi​​f(v,⋅)−infZi​​f(v,⋅)≤d(2ϵ(n)).
  4. Eq. (7): with Mn=C∫∣hn−h∗∣ dxM_n=C\int|h_n-h^*|\,dxMn​=C∫∣hn​−h∗∣dx, for every vvv,
Jn(v)−Mn≤∫f(v,x)h∗(x) dx≤Jn(v)+Mn+d(2ϵ(n)).J_n(v)-M_n\le\int f(v,x)h^*(x)\,dx\le J_n(v)+M_n+d(2\epsilon(n)).Jn​(v)−Mn​≤∫f(v,x)h∗(x)dx≤Jn​(v)+Mn​+d(2ϵ(n)).
  1. Strong L1L^1L1 consistency of the box kernel density estimator: if ϵ(n)→0\epsilon(n)\to0ϵ(n)→0 and nϵ(n)m→∞n\epsilon(n)^m\to\inftynϵ(n)m→∞, then ∫∣hn−h∗∣ dx→0\int|h_n-h^*|\,dx\to0∫∣hn​−h∗∣dx→0 almost surely.

Milestones 1–4 are deterministic statements about a fixed sample; milestone 5 is the only probabilistic input.

Significance

Theorem 3.1 gives consistency of a tractable robust reformulation of a sampled stochastic program under conditions the paper notes are weaker than those of King and Wets for sampled stochastic programs: fff need only be bounded and equicontinuous in xxx, uniformly in vvv, and the true distribution need only have a density. It also gives an explicit schedule for the size of the uncertainty set, ϵ(n)→0\epsilon(n)\to0ϵ(n)→0 with nϵ(n)m→∞n\epsilon(n)^m\to\inftynϵ(n)m→∞, the bandwidth condition of kernel density estimation. Section 4 of the paper applies the same distributional interpretation to regularised learning methods such as the support vector machine and the Lasso.

The result is proved in the paper, with the L1L^1L1 consistency of kernel density estimators (Devroye 1983; Devroye and Györfi 1985) cited rather than proved. No part of it is formalized in Lean or on this platform as far as a search of the platform found. A complete development would produce, besides Theorem 3.1, a machine-checked strong L1L^1L1 consistency theorem for kernel density estimators, which is a basic result of nonparametric statistics in its own right.

Difficulty

The deterministic part (milestones 1–4) is measure-theoretic bookkeeping: the kernel integrates to one only because the box is a sup-norm ball of volume (2ϵ)m(2\epsilon)^m(2ϵ)m, and every infimum and supremum must be handled with care, since fff need not attain them.

The obstacle is milestone 5. Almost-sure L1L^1L1 convergence of hnh_nhn​ to an arbitrary density h∗h^*h∗, with no continuity or support assumption, does not follow from the strong law of large numbers applied pointwise: hn(x)h_n(x)hn​(x) is an average of nnn terms whose law changes with nnn through ϵ(n)\epsilon(n)ϵ(n), and almost-sure convergence at each fixed xxx does not give convergence of the integral along a single sample path. The theorem needs both a bias estimate valid for every integrable density and a concentration estimate for the random L1L^1L1 error. Mathlib has Lebesgue differentiation and the strong law, but no kernel density estimator and no such concentration result.

Formalization scope

  • Rm\mathbb R^mRm is Fin m → ℝ, whose Mathlib norm is the sup norm; boxes are Metric.closedBall. The integrals ∫f(v,x)h∗(x) dx\int f(v,x)h^*(x)\,dx∫f(v,x)h∗(x)dx are Bochner integrals against Lebesgue measure of integrable integrands.
  • The samples are a sequence X : ℕ → Ω → Fin m → ℝ on a probability space, independent (iIndepFun) and each with law volume.withDensity h*; x1,x2,…x_1,x_2,\dotsx1​,x2​,… become X 0, X 1, …, and the nnn-th problem uses the first nnn. "With probability one" is ∀ᵐ ω ∂P.
  • The goal quantifies over every selection v(n)v(n)v(n) of maximisers, with no measurability assumed; a version with one chosen maximiser would be weaker and is ruled out.
  • Readings and corrections of the printed text:
    • the kernel argument printed (x−xi)/ϵ(x-x_i)/\epsilon(x−xi​)/ϵ on p. 98 is read as (x−xi)/ϵ(n)(x-x_i)/\epsilon(n)(x−xi​)/ϵ(n), as the proof on p. 99 writes it;
    • "max⁡v,x∣f(v,x)∣≤C\max_{v,x}|f(v,x)|\le Cmaxv,x​∣f(v,x)∣≤C" is read as the uniform bound ∣f∣≤C|f|\le C∣f∣≤C and the "max" in d(ϵ)d(\epsilon)d(ϵ) as a supremum;
    • "d(ϵ)↓0d(\epsilon)\downarrow0d(ϵ)↓0" is read as d(ϵ)→0d(\epsilon)\to0d(ϵ)→0 as ϵ↓0\epsilon\downarrow0ϵ↓0;
    • implicit hypotheses made explicit: F≠∅\mathcal F\ne\emptysetF=∅, ϵ(n)>0\epsilon(n)>0ϵ(n)>0, measurability of f(v,⋅)f(v,\cdot)f(v,⋅), h∗h^*h∗ a Lebesgue density;
    • the monotonicity in "ϵ(n)↓0\epsilon(n)\downarrow0ϵ(n)↓0, nϵ(n)m↑∞n\epsilon(n)^m\uparrow\inftynϵ(n)m↑∞" is kept in the goal; milestone 5 uses the limits only, as the paper states it;
    • the paper's MnM_nMn​ ("there exists {Mn}→0\{M_n\}\to0{Mn​}→0") is made explicit as Mn=C∫∣hn−h∗∣M_n=C\int|h_n-h^*|Mn​=C∫∣hn​−h∗∣, so Eq. (7) is stated for every sample.
  • Remark 3.2 and Appendix B (an integrable envelope in place of boundedness) are not part of this mission.
  • Every real infimum and supremum ranges over a nonempty set of values bounded by CCC in absolute value, so no statement holds through a junk value; a formalization in which the supremum over F\mathcal FF or the box infimum could be vacuous is excluded.
  • The definitions (boxes, Pn\mathcal P_nPn​, the kernel, the estimator, JnJ_nJn​, ddd) live in one definition file. Pn\mathcal P_nPn​ duplicates, with weights 1/n1/n1/n, the distribution set of mission I; the duplication is deliberate because draft missions cannot import each other.
  • Welcome contributions: the kernel density estimator and its strong L1L^1L1 consistency as reusable infrastructure, and any of the deterministic milestones.

Selected references

  • H. Xu, C. Caramanis, S. Mannor, A Distributional Interpretation of Robust Optimization, Mathematics of Operations Research 37(1):95–110, 2012. https://doi.org/10.1287/moor.1110.0531
  • L. Devroye, The equivalence of weak, strong and complete convergence in L1L_1L1​ for kernel density estimates, Annals of Statistics 11(3):896–904, 1983.
  • L. Devroye, L. Györfi, Nonparametric Density Estimation: The L1L_1L1​ View, Wiley, 1985.
  • A. J. King, R. J.-B. Wets, Epi-consistency of convex stochastic programs, Stochastics and Stochastic Reports 34(1), 1991 (reference [22] of the paper).
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ProbabilityStatistics·Captain: mikedeng1

Learnability, Stability and Uniform Convergence III: For an ERM, Leave-One-Out Stability, Universal Consistency and Universal Generalization Are EquivalentResearch Paper

Motivation

Algorithmic stability asks how much the output of a learning algorithm changes when its training sample is perturbed. Since Devroye and Wagner (IEEE Trans. Inf. Theory 1979) it has served as a route to generalization bounds that does not go through the complexity of the hypothesis class. Bousquet and Elisseeff (JMLR 2002) popularised uniform stability, and Mukherjee, Niyogi, Poggio and Rifkin (Adv. Comput. Math. 2006) showed that for empirical risk minimisation in supervised learning, a leave-one-out type of stability is necessary and sufficient for consistency.

Shalev-Shwartz, Shamir, Srebro and Sridharan (JMLR 11, 2010) study stability in Vapnik's General Learning Setting, where uniform convergence can fail even though the problem is learnable. In Appendix A.2 they compare replace-one and leave-one-out (LOO) stability. For an empirical risk minimiser they prove that LOO stability is equivalent to consistency and to generalization, provided each property holds with one rate for all distributions (Theorem 31, p. 2667). This mission formalizes that theorem and the lemmas of Section 5.3 on which its proof rests.

Timeline:

  • 1979, Devroye–Wagner: leave-one-out estimates for local rules.
  • 2002, Bousquet–Elisseeff: uniform stability implies generalization.
  • 2002, Kutin–Niyogi (UAI 2002): a taxonomy of stability notions.
  • 2006, Mukherjee et al.: LOO stability characterises consistency of ERM in supervised learning.
  • 2010, Shalev-Shwartz et al.: in the General Learning Setting, for ERMs, LOO stability, universal consistency and universal generalization are equivalent (Theorem 31). Universally consistent AERMs need not be LOO stable (Example 6).

Setting

A learning problem consists of an instance space Z\mathcal ZZ with a σ\sigmaσ-algebra, a nonempty hypothesis class H\mathcal HH, and an objective f:H×Z→Rf:\mathcal H\times\mathcal Z\to\mathbb Rf:H×Z→R with ∣f(h;z)∣≤B|f(h;z)|\le B∣f(h;z)∣≤B for all h,zh,zh,z. For a probability measure D\mathcal DD on Z\mathcal ZZ:

  • the risk is F(h)=Ez∼D[f(h;z)]F(h)=\mathbb E_{z\sim\mathcal D}[f(h;z)]F(h)=Ez∼D​[f(h;z)] and the optimal risk is F∗=inf⁡hF(h)F^*=\inf_{h}F(h)F∗=infh​F(h);
  • for a sample S=(z1,…,zm)∼DmS=(z_1,\dots,z_m)\sim\mathcal D^mS=(z1​,…,zm​)∼Dm of mmm i.i.d. draws, the empirical risk is FS(h)=1m∑if(h;zi)F_S(h)=\frac1m\sum_{i}f(h;z_i)FS​(h)=m1​∑i​f(h;zi​), and FS(h^S)=inf⁡hFS(h)F_S(\hat h_S)=\inf_hF_S(h)FS​(h^S​)=infh​FS​(h) denotes the minimal empirical risk;
  • a learning rule AAA maps each sample of size m≥1m\ge1m≥1 to a hypothesis A(S)A(S)A(S). It is an ERM if FS(A(S))=FS(h^S)F_S(A(S))=F_S(\hat h_S)FS​(A(S))=FS​(h^S​) for every sample. It is an AERM with rate εerm\varepsilon_{\mathrm{erm}}εerm​ if E[FS(A(S))−FS(h^S)]≤εerm(m)\mathbb E[F_S(A(S))-F_S(\hat h_S)]\le\varepsilon_{\mathrm{erm}}(m)E[FS​(A(S))−FS​(h^S​)]≤εerm​(m);
  • AAA is consistent with rate ε\varepsilonε if ES∼Dm[F(A(S))−F∗]≤ε(m)\mathbb E_{S\sim\mathcal D^m}[F(A(S))-F^*]\le\varepsilon(m)ES∼Dm​[F(A(S))−F∗]≤ε(m). It generalizes with rate ε\varepsilonε if E[∣F(A(S))−FS(A(S))∣]≤ε(m)\mathbb E[|F(A(S))-F_S(A(S))|]\le\varepsilon(m)E[∣F(A(S))−FS​(A(S))∣]≤ε(m), and it on-average generalizes if ∣E[F(A(S))−FS(A(S))]∣≤ε(m)|\mathbb E[F(A(S))-F_S(A(S))]|\le\varepsilon(m)∣E[F(A(S))−FS​(A(S))]∣≤ε(m);
  • writing S∖iS^{\setminus i}S∖i for SSS with ziz_izi​ removed, AAA is LOO stable with rate ε\varepsilonε (Definition 29) if
1m∑i=1mES∼Dm[∣f(A(S∖i);zi)−f(A(S);zi)∣]≤ε(m).\frac1m\sum_{i=1}^m\mathbb E_{S\sim\mathcal D^m}\Big[\big|f(A(S^{\setminus i});z_i)-f(A(S);z_i)\big|\Big]\le\varepsilon(m).m1​i=1∑m​ES∼Dm​[​f(A(S∖i);zi​)−f(A(S);zi​)​]≤ε(m).

A rate is a sequence ε(m)\varepsilon(m)ε(m) that is non-increasing and tends to 000. A property holds universally if it holds under every D\mathcal DD with one and the same rate.

Formalization targets

Goal: Theorem 31

For an ERM AAA:

A universally LOO stable  ⟺  A universally consistent  ⟺  A universally generalizes.A\ \text{universally LOO stable}\iff A\ \text{universally consistent}\iff A\ \text{universally generalizes}.A universally LOO stable⟺A universally consistent⟺A universally generalizes.

The statement has no rates. Each side asserts that some rate exists and serves every distribution.

Milestones

In the order the proof uses them:

  1. Utility Lemma 12: E∣X−EX∣≤B/m\mathbb E|X-\mathbb EX|\le B/\sqrt mE∣X−EX∣≤B/m​ for the mean XXX of mmm i.i.d. variables bounded by BBB.
  2. Utility Lemma 13: X≤YX\le YX≤Y a.s. implies E∣X∣≤∣EX∣+2E∣Y∣\mathbb E|X|\le|\mathbb EX|+2\mathbb E|Y|E∣X∣≤∣EX∣+2E∣Y∣.
  3. Lemma 14: an AERM that on-average generalizes with rate εoag\varepsilon_{\mathrm{oag}}εoag​ generalizes with rate εoag+2εerm+2B/m\varepsilon_{\mathrm{oag}}+2\varepsilon_{\mathrm{erm}}+2B/\sqrt mεoag​+2εerm​+2B/m​.
  4. Lemma 15: under the same hypotheses the rule is consistent with rate εoag+εerm\varepsilon_{\mathrm{oag}}+\varepsilon_{\mathrm{erm}}εoag​+εerm​.
  5. Lemma 16 (Main Converse Lemma): in a learnable problem, E∣FS(h^S)−F∗∣≤2εcons(m′)+2B/m+2Bm′2/m\mathbb E|F_S(\hat h_S)-F^*|\le2\varepsilon_{\mathrm{cons}}(m')+2B/\sqrt m+2Bm'^2/mE∣FS​(h^S​)−F∗∣≤2εcons​(m′)+2B/m​+2Bm′2/m for 2≤m′≤m/22\le m'\le m/22≤m′≤m/2.
  6. Lemma 17: Eq. (12), together with an AERM that is consistent, gives generalization with rate εemp+εerm+εcons\varepsilon_{\mathrm{emp}}+\varepsilon_{\mathrm{erm}}+\varepsilon_{\mathrm{cons}}εemp​+εerm​+εcons​.
  7. First display of the proof of Theorem 31: a generalizing ERM is LOO stable with rate εgen(m−1)\varepsilon_{\mathrm{gen}}(m-1)εgen​(m−1).
  8. Second display: a LOO stable ERM on-average generalizes on samples of size m−1m-1m−1 with rate εstable(m)+2B/m\varepsilon_{\mathrm{stable}}(m)+2B/mεstable​(m)+2B/m.

Significance

Theorem 31 shows that for exact ERMs, LOO stability is not only sufficient but necessary for consistency. It transfers the supervised-learning characterisation of Mukherjee et al. to the General Learning Setting, where uniform convergence is no longer available as an intermediate. The hypothesis is sharp in one direction: Example 6 of the paper gives a universally consistent AERM that is not LOO stable. The equivalence therefore depends on exact minimisation, and an asymptotic minimiser does not suffice.

The lemmas are useful on their own. Lemmas 14 and 15 are the standard bridges between on-average generalization, generalization and consistency. Lemma 16 says that the minimal empirical risk estimates F∗F^*F∗ consistently in every learnable problem, even when no ERM learns. Lemma 16 also underlies Theorem 7, the paper's main characterisation of learnability, which is the subject of mission I of this series.

The results are proved in the paper. As far as is known they have not been formalized in any proof assistant. The platform has the textbook side of this framework (Shalev-Shwartz and Ben-David, Understanding Machine Learning, Chapter 13). Those statements are in Rd\mathbb R^dRd and use a replace-one stability notion; they do not cover leave-one-out stability or the General Learning Setting.

Difficulty

The implications between stability and generalization change the sample size: S∖iS^{\setminus i}S∖i has m−1m-1m−1 points, so a statement about the rule at size mmm has to be compared with the rule at size m−1m-1m−1 under the marginal law of the reduced sample. The step from universal consistency to generalization needs Lemma 16. That lemma estimates F∗F^*F∗ from a sample on which the ERM itself may be inconsistent, and it is the only place where universality of the consistency rate is used. Per-distribution consistency of an ERM does not imply generalization (Example 1 of the paper). An argument that fixes D\mathcal DD throughout therefore cannot succeed.

Formalization scope

Samples are tuples S:Fin m→ZS:\mathrm{Fin}\,m\to\mathcal ZS:Finm→Z with law Measure.pi (the i.i.d. product), and S∖iS^{\setminus i}S∖i is Fin.removeNth i S. A learning rule is a family Am:Zm→HA_m:\mathcal Z^m\to\mathcal HAm​:Zm→H. Its value at m=0m=0m=0 is never used, and LOO stability is required only for m≥2m\ge2m≥2. FS(h^S)F_S(\hat h_S)FS​(h^S​) is the infimum inf⁡hFS(h)\inf_hF_S(h)infh​FS​(h) and no minimiser is chosen. An ERM is a rule attaining this infimum at every sample. A rate is non-increasing on m≥1m\ge1m≥1 and tends to 000. Universal properties are stated as "there exists ε\varepsilonε with IsRate ε such that for every probability measure D\mathcal DD …", with the rate chosen before the distribution. The bound BBB is any bound on ∣f∣|f|∣f∣; the paper's BBB is sup⁡∣f∣\sup|f|sup∣f∣, and all its rates increase with BBB.

Measurability is not discussed in the paper. The formalization assumes that each f(h;⋅)f(h;\cdot)f(h;⋅) is measurable and that the rule is measurable in the sense that (S,z)↦f(Am(S);z)(S,z)\mapsto f(A_m(S);z)(S,z)↦f(Am​(S);z) is jointly measurable. Lemmas 14–17 also assume that S↦inf⁡hFS(h)S\mapsto\inf_hF_S(h)S↦infh​FS​(h) is measurable. For a measurable ERM this holds automatically, so Theorem 31 makes no such assumption. Without these assumptions Lean's integral of a non-measurable function is 000 and every rate bound would hold trivially. For the same reason Utility Lemma 13 assumes X,YX,YX,Y integrable. The ERM hypothesis of the goal must not be weakened to an AERM: the statement would then be false (Example 6).

One statement corrects the printed text. In the second display of the proof of Theorem 31 (p. 2668), the chain adds 2B/m2B/m2B/m and then drops it. The milestone states the bound the argument proves, εstable(m)+2B/m\varepsilon_{\mathrm{stable}}(m)+2B/mεstable​(m)+2B/m. The rate-free Theorem 31 is unaffected.

The development needs product measures, independence and variance bounds, all available in Mathlib, and the marginals of Measure.pi under removal of a coordinate. The definitions of risks, rules and stability notions can be reused by other stability results. Proofs of any milestone are welcome, and so are alternative proofs of the goal.

Selected references

  • S. Shalev-Shwartz, O. Shamir, N. Srebro, K. Sridharan, Learnability, Stability and Uniform Convergence, Journal of Machine Learning Research 11 (2010) 2635–2670. https://jmlr.org/papers/v11/shalev-shwartz10a.html
  • O. Bousquet, A. Elisseeff, Stability and Generalization, Journal of Machine Learning Research 2 (2002) 499–526. https://www.jmlr.org/papers/v2/bousquet02a.html
  • S. Mukherjee, P. Niyogi, T. Poggio, R. Rifkin, Learning theory: stability is sufficient for generalization and necessary and sufficient for consistency of empirical risk minimization, Advances in Computational Mathematics 25 (2006) 161–193. https://doi.org/10.1007/s10444-004-7634-z
  • S. Kutin, P. Niyogi, Almost-everywhere algorithmic stability and generalization error, UAI 2002. https://arxiv.org/abs/1301.0579
  • L. Devroye, T. Wagner, Distribution-free performance bounds for potential function rules, IEEE Transactions on Information Theory 25(5) (1979) 601–604. https://doi.org/10.1109/TIT.1979.1056087
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Convex OptimizationOptimizationProbability+1·Captain: mikedeng1

Learnability, Stability and Uniform Convergence II: Tikhonov-Regularized ERM Learns Convex Lipschitz Stochastic Optimization in Hilbert Space with High ProbabilityResearch Paper

Motivation

Statistical learning theory asks when a rule that sees only an i.i.d. sample z1,…,zmz_1,\dots,z_mz1​,…,zm​ from an unknown distribution DDD can return a hypothesis whose expected loss is close to the best possible. In supervised classification the classical answer is uniform convergence: learnability holds exactly when empirical risks converge to expected risks uniformly over the hypothesis class, and then empirical risk minimization (ERM) learns. Shalev-Shwartz, Shamir, Srebro and Sridharan (JMLR 11, 2010) showed that in Vapnik's broader General Learning Setting this picture breaks down. Their motivating example is stochastic convex optimization in a Hilbert space: minimizing an expected convex, Lipschitz objective over a bounded convex set from samples. This problem underlies regularized linear prediction, kernel methods and online-to-batch conversions, and the paper shows (§4.1) that in infinite dimension uniform convergence can fail and the plain empirical minimizer can fail to converge, while the problem is still learnable.

This mission formalizes the positive half of that example: Tikhonov-regularized ERM learns every such problem, with an explicit bound holding with probability 1−δ1-\delta1−δ (Theorem 3, p. 2644), through the stability of strongly convex empirical minimization (Theorem 2).

Setting

Let ZZZ be a measurable space of instances and EEE a real Hilbert space. A stochastic convex optimization problem consists of a nonempty, closed, convex, bounded set H⊆E\mathcal H\subseteq EH⊆E and an objective f:E×Z→Rf:E\times Z\to\mathbb Rf:E×Z→R such that for every zzz the map h↦f(h;z)h\mapsto f(h;z)h↦f(h;z) is convex and LLL-Lipschitz on H\mathcal HH, each f(h;⋅)f(h;\cdot)f(h;⋅) is measurable, and ∣f(h;z)∣≤C|f(h;z)|\le C∣f(h;z)∣≤C on H×Z\mathcal H\times ZH×Z. For a distribution DDD on ZZZ define the risk and optimal risk

F(h)=Ez∼D[f(h;z)],F∗=inf⁡h∈HF(h),F(h)=\mathbb E_{z\sim D}[f(h;z)],\qquad F^*=\inf_{h\in\mathcal H}F(h),F(h)=Ez∼D​[f(h;z)],F∗=h∈Hinf​F(h),

and for a sample S=(z1,…,zm)∼DmS=(z_1,\dots,z_m)\sim D^mS=(z1​,…,zm​)∼Dm the empirical risk FS(h)=1m∑i=1mf(h;zi)F_S(h)=\frac1m\sum_{i=1}^m f(h;z_i)FS​(h)=m1​∑i=1m​f(h;zi​). A function ggg is λ\lambdaλ-strongly convex on H\mathcal HH if g−λ2∥⋅∥2g-\frac\lambda2\|\cdot\|^2g−2λ​∥⋅∥2 is convex there. The regularized empirical minimizer is

h^λ∈arg min⁡h∈H(FS(h)+λ2∥h∥2).(5)\hat h_\lambda\in\operatorname*{arg\,min}_{h\in\mathcal H}\Big(F_S(h)+\frac\lambda2\|h\|^2\Big).\tag{5}h^λ​∈h∈Hargmin​(FS​(h)+2λ​∥h∥2).(5)

For the general part, a learning rule AAA maps samples to hypotheses; it is an AERM with rate εerm\varepsilon_{\mathrm{erm}}εerm​ if E[FS(A(S))−inf⁡hFS(h)]≤εerm(m)\mathbb E[F_S(A(S))-\inf_hF_S(h)]\le\varepsilon_{\mathrm{erm}}(m)E[FS​(A(S))−infh​FS​(h)]≤εerm​(m), consistent with rate εcons\varepsilon_{\mathrm{cons}}εcons​ if E[F(A(S))−F∗]≤εcons(m)\mathbb E[F(A(S))-F^*]\le\varepsilon_{\mathrm{cons}}(m)E[F(A(S))−F∗]≤εcons​(m), and uniform-RO stable with rate εstable\varepsilon_{\mathrm{stable}}εstable​ if replacing any one sample point changes the loss at any test point by at most εstable(m)\varepsilon_{\mathrm{stable}}(m)εstable​(m) on average over the replaced index (Definition 4).

Formalization targets

Goal: Theorem 3

If ∥h∥≤B\|h\|\le B∥h∥≤B on H\mathcal HH, L,B>0L,B>0L,B>0, δ∈(0,1)\delta\in(0,1)δ∈(0,1), m≥1m\ge1m≥1 and λ=16L2/(δB2m)\lambda=\sqrt{16L^2/(\delta B^2m)}λ=16L2/(δB2m)​, then with probability at least 1−δ1-\delta1−δ over S∼DmS\sim D^mS∼Dm

F(h^λ)−F∗ ≤ 4L2B2δm(1+8δm).F(\hat h_\lambda)-F^*\ \le\ 4\sqrt{\frac{L^2B^2}{\delta m}}\Big(1+\frac8{\delta m}\Big).F(h^λ​)−F∗ ≤ 4δmL2B2​​(1+δm8​).

The constants are the paper's.

Milestones, in the order the proof uses them

  1. Quadratic growth at a minimizer of a λ\lambdaλ-strongly convex ggg: g(h′)−g(h)≥λ2∥h′−h∥2g(h')-g(h)\ge\frac\lambda2\|h'-h\|^2g(h′)−g(h)≥2λ​∥h′−h∥2 (§4.2, p. 2644).
  2. Eq. (6): if f(⋅;z)f(\cdot;z)f(⋅;z) is λ\lambdaλ-strongly convex and LLL-Lipschitz, empirical minimizers of SSS and of S(i)S^{(i)}S(i) satisfy ∣f(h^S,z)−f(h^S(i),z)∣≤4L2/(λm)|f(\hat h_S,z)-f(\hat h_S^{(i)},z)|\le 4L^2/(\lambda m)∣f(h^S​,z)−f(h^S(i)​,z)∣≤4L2/(λm) for all zzz (p. 2645).
  3. Theorem 8: a uniform- or average-RO stable AERM is consistent with rate εstable+εerm\varepsilon_{\mathrm{stable}}+\varepsilon_{\mathrm{erm}}εstable​+εerm​ and generalizes with rate εstable+2εerm+2C/m\varepsilon_{\mathrm{stable}}+2\varepsilon_{\mathrm{erm}}+2C/\sqrt mεstable​+2εerm​+2C/m​ (p. 2649).
  4. ES∼Dm[F(h^S)−F∗]≤4L2/(λm)\mathbb E_{S\sim D^m}[F(\hat h_S)-F^*]\le 4L^2/(\lambda m)ES∼Dm​[F(h^S​)−F∗]≤4L2/(λm) for the strongly convex empirical minimizer (p. 2645).
  5. Theorem 2: with probability 1−δ1-\delta1−δ, F(h^S)−F∗≤4L2/(δλm)F(\hat h_S)-F^*\le 4L^2/(\delta\lambda m)F(h^S​)−F∗≤4L2/(δλm) (p. 2644).
  6. Theorem 2 applied to r(h;z)=λ2∥h∥2+f(h;z)r(h;z)=\frac\lambda2\|h\|^2+f(h;z)r(h;z)=2λ​∥h∥2+f(h;z): with probability 1−δ1-\delta1−δ, λ2∥h^λ∥2+F(h^λ)≤inf⁡h(λ2∥h∥2+F(h))+4(L+λB)2/(δλm)\frac\lambda2\|\hat h_\lambda\|^2+F(\hat h_\lambda)\le\inf_h\big(\frac\lambda2\|h\|^2+F(h)\big)+4(L+\lambda B)^2/(\delta\lambda m)2λ​∥h^λ​∥2+F(h^λ​)≤infh​(2λ​∥h∥2+F(h))+4(L+λB)2/(δλm) (p. 2645).

Significance

The result. Theorem 3 shows that every convex, Lipschitz, bounded stochastic optimization problem over a bounded subset of a Hilbert space is learnable at rate O(LB/δm)O(LB/\sqrt{\delta m})O(LB/δm​), with no dimension dependence and no uniform convergence. Together with the counterexamples of §4.1 it separates learnability from uniform convergence and from ERM, and it motivates the paper's general characterization: a problem is learnable if and only if it admits a uniform-RO stable asymptotic empirical risk minimizer (Theorem 7). Theorem 8 is the sufficiency half of that characterization and is reused wherever stability arguments give generalization bounds.

Formalizing it. The results are proved in the paper; to our knowledge none has a machine-checked proof. The closest platform material is the textbook treatment in Understanding Machine Learning, chapter 13 (Shalev-Shwartz and Ben-David): Corollary 13.9 (UnderstandingML.convex_lipschitz_bounded_learnable), Corollary 13.6 (rlm_lipschitz_stable) and Lemma 13.5 (strongly_convex_lemma). Those are stated in Rd\mathbb R^dRd, bound the risk in expectation, use the regularizer λ∥w∥2\lambda\|w\|^2λ∥w∥2 over all of Rd\mathbb R^dRd, and have different constants; the present mission works in an arbitrary Hilbert space, over a constraint set H\mathcal HH, with high-probability bounds and the paper's constants. Its definitions of learning rules, AERM, consistency and replace-one stability in the General Learning Setting are reusable by the other missions of this series.

Difficulty

The obvious route, bounding sup⁡h∈H∣F(h)−FS(h)∣\sup_{h\in\mathcal H}|F(h)-F_S(h)|suph∈H​∣F(h)−FS​(h)∣, is unavailable: §4.1 exhibits problems of exactly this type in which that supremum stays bounded away from zero for every sample size. Any successful argument therefore has to rely on a property of the learning rule rather than of the class H\mathcal HH, and the plain empirical minimizer does not have it: §4.1 shows it can stay a constant away from F∗F^*F∗ at every sample size. A second difficulty is purely formal: the regularization parameter λ\lambdaλ depends on δ\deltaδ and mmm, so the regularized minimizer changes with them, and all expectations involve a data-dependent hypothesis in a possibly non-separable Hilbert space, where measurability is not automatic.

Formalization scope

Lean conventions, fixed for every item:

  • EEE is a real inner product space with CompleteSpace E, never assumed finite-dimensional; H\mathcal HH is Hset : Set E, and all infima, suprema, strong convexity and Lipschitz conditions are taken on Hset only. F∗F^*F∗ is ⨅ h : Hset, F h.
  • Samples are Fin m → Z, DmD^mDm is Measure.pi, S(i)S^{(i)}S(i) is Function.update S i z', and m≥1m\ge1m≥1 throughout.
  • The paper's standing loss bound ∣f∣≤B|f|\le B∣f∣≤B (p. 2637) is named CCC, because Theorem 3 uses BBB for the norm bound ∥h∥≤B\|h\|\le B∥h∥≤B. L>0L>0L>0 and B>0B>0B>0 are implicit in Theorem 3's choice of λ\lambdaλ and are stated.
  • Strong convexity is Mathlib's StrongConvexOn Hset λ, which is the paper's definition.
  • Minimizers are selections S↦h^S∈HS\mapsto\hat h_S\in\mathcal HS↦h^S​∈H satisfying the minimization property; the theorems hold for every such selection, hence for the minimizer, which is unique by strong convexity.
  • Measurability, not discussed in the paper, is the series' single standing convention: each f(h;⋅)f(h;\cdot)f(h;⋅) is measurable and the selection makes (S,z)↦f(h^S;z)(S,z)\mapsto f(\hat h_S;z)(S,z)↦f(h^S​;z) jointly measurable; for Theorem 8, the rule is measurable in the same sense and S↦inf⁡hFS(h)S\mapsto\inf_hF_S(h)S↦infh​FS​(h) is measurable.
  • "With probability at least 1−δ1-\delta1−δ" is the bound Dm{failure}≤δD^m\{\text{failure}\}\le\deltaDm{failure}≤δ with 0<δ<10<\delta<10<δ<1.

No statement of the paper is corrected: all printed constants were checked against the proofs and are reproduced exactly.

A formalization in which the expected excess risk is a Bochner integral of a non-measurable or non-integrable function, or in which F∗F^*F∗ is an infimum over all of EEE or over an unbounded family, would make the bounds trivially true; the measurability hypotheses, the bound ∣f∣≤C|f|\le C∣f∣≤C and the infimum over the nonempty set H\mathcal HH rule this out.

Contributions welcome: proofs of the milestones in order, and in particular a reusable replace-one identity E[FS(A(S))]=1m∑iE[f(A(S(i));zi′)]\mathbb E[F_S(A(S))]=\frac1m\sum_i\mathbb E[f(A(S^{(i)});z'_i)]E[FS​(A(S))]=m1​∑i​E[f(A(S(i));zi′​)] under Measure.pi, and Markov's inequality in the form used for high-probability bounds.

Selected references

  • S. Shalev-Shwartz, O. Shamir, N. Srebro, K. Sridharan, Learnability, Stability and Uniform Convergence, Journal of Machine Learning Research 11 (2010) 2635–2670. https://jmlr.org/papers/v11/shalev-shwartz10a.html
  • S. Shalev-Shwartz, O. Shamir, N. Srebro, K. Sridharan, Stochastic Convex Optimization, COLT 2009. https://www.cs.mcgill.ca/~colt2009/papers/018.pdf
  • O. Bousquet, A. Elisseeff, Stability and Generalization, Journal of Machine Learning Research 2 (2002) 499–526. https://jmlr.org/papers/v2/bousquet02a.html
  • S. Shalev-Shwartz, S. Ben-David, Understanding Machine Learning: From Theory to Algorithms, Cambridge University Press, 2014, chapter 13. https://doi.org/10.1017/CBO9781107298019
  • V. N. Vapnik, Statistical Learning Theory, Wiley, 1998.
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Convex OptimizationOptimizationStatistics·Captain: mikedeng1

A Unified Framework for High-Dimensional Analysis of M-Estimators with Decomposable Regularizers: Error Bounds under Decomposability and Restricted Strong ConvexityResearch Paper

Motivation

High-dimensional statistics studies estimation when the number of parameters ppp is comparable to, or larger than, the number of observations nnn. The standard estimators in this regime are regularized M-estimators: minimise an empirical loss plus a penalty that encodes structure, such as the Lasso (ℓ1\ell_1ℓ1​ penalty, sparse vectors), the group Lasso (block norms, group sparsity) and nuclear-norm regularization (low-rank matrices). Before 2009 each of these estimators came with its own consistency proof. Negahban, Ravikumar, Wainwright and Yu (arXiv:1010.2731; Statistical Science 27(4), 2012, doi:10.1214/12-STS400) isolated two properties that these proofs share, decomposability of the regularizer and restricted strong convexity of the loss, and proved one deterministic theorem from them. The Lasso rates of Bickel, Ritov and Tsybakov (arXiv:0801.1095), rates under ℓq\ell_qℓq​-sparsity, and group-sparse and low-rank rates then follow as corollaries. The framework is the organising principle of Chapter 9 of Wainwright's textbook High-Dimensional Statistics (Cambridge University Press, 2019).

Setting

Let EEE be a finite-dimensional real inner product space with inner product ⟨⋅,⋅⟩\langle\cdot,\cdot\rangle⟨⋅,⋅⟩ and induced error norm ∥⋅∥\|\cdot\|∥⋅∥. Given a loss L:E→R\mathcal L:E\to\mathbb RL:E→R, a regularizer R:E→R\mathcal R:E\to\mathbb RR:E→R and a constant λn>0\lambda_n>0λn​>0, program (1) is

θ^λn∈arg⁡min⁡θ∈E{L(θ)+λnR(θ)}.\hat\theta_{\lambda_n}\in\arg\min_{\theta\in E}\{\mathcal L(\theta)+\lambda_n\mathcal R(\theta)\}.θ^λn​​∈argθ∈Emin​{L(θ)+λn​R(θ)}.

For a subspace SSS write uSu_SuS​ for the orthogonal projection of uuu onto SSS, and S⊥S^\perpS⊥ for the orthogonal complement.

  • Decomposability. For subspaces M⊆M‾\mathcal M\subseteq\overline{\mathcal M}M⊆M, the norm R\mathcal RR is decomposable with respect to (M,M‾⊥)(\mathcal M,\overline{\mathcal M}^\perp)(M,M⊥) if R(θ+γ)=R(θ)+R(γ)\mathcal R(\theta+\gamma)=\mathcal R(\theta)+\mathcal R(\gamma)R(θ+γ)=R(θ)+R(γ) for all θ∈M\theta\in\mathcal Mθ∈M and γ∈M‾⊥\gamma\in\overline{\mathcal M}^\perpγ∈M⊥. Example: the ℓ1\ell_1ℓ1​-norm with M=M‾={θ:θj=0 ∀j∉S}\mathcal M=\overline{\mathcal M}=\{\theta:\theta_j=0\ \forall j\notin S\}M=M={θ:θj​=0 ∀j∈/S}.
  • Dual norm. R∗(v)=sup⁡R(u)≤1⟨u,v⟩\mathcal R^*(v)=\sup_{\mathcal R(u)\le1}\langle u,v\rangleR∗(v)=supR(u)≤1​⟨u,v⟩.
  • Subspace compatibility constant. Ψ(M‾)=sup⁡u∈M‾∖{0}R(u)/∥u∥\Psi(\overline{\mathcal M})=\sup_{u\in\overline{\mathcal M}\setminus\{0\}}\mathcal R(u)/\|u\|Ψ(M)=supu∈M∖{0}​R(u)/∥u∥; for the ℓ1\ell_1ℓ1​-norm on an sss-dimensional coordinate subspace, Ψ=s\Psi=\sqrt sΨ=s​.
  • The set C\mathbb CC. For a point θ∗∈E\theta^*\in Eθ∗∈E,
C(M,M‾⊥;θ∗)={Δ∣R(ΔM‾⊥)≤3R(ΔM‾)+4R(θM⊥∗)}.\mathbb C(\mathcal M,\overline{\mathcal M}^\perp;\theta^*)=\{\Delta\mid\mathcal R(\Delta_{\overline{\mathcal M}^\perp})\le3\mathcal R(\Delta_{\overline{\mathcal M}})+4\mathcal R(\theta^*_{\mathcal M^\perp})\}.C(M,M⊥;θ∗)={Δ∣R(ΔM⊥​)≤3R(ΔM​)+4R(θM⊥∗​)}.
  • Restricted strong convexity (RSC). With the Taylor error δL(Δ,θ∗)=L(θ∗+Δ)−L(θ∗)−⟨∇L(θ∗),Δ⟩\delta\mathcal L(\Delta,\theta^*)=\mathcal L(\theta^*+\Delta)-\mathcal L(\theta^*)-\langle\nabla\mathcal L(\theta^*),\Delta\rangleδL(Δ,θ∗)=L(θ∗+Δ)−L(θ∗)−⟨∇L(θ∗),Δ⟩, the loss satisfies RSC with curvature κL>0\kappa_{\mathcal L}>0κL​>0 and tolerance τL(θ∗)\tau_{\mathcal L}(\theta^*)τL​(θ∗) if δL(Δ,θ∗)≥κL∥Δ∥2−τL2(θ∗)\delta\mathcal L(\Delta,\theta^*)\ge\kappa_{\mathcal L}\|\Delta\|^2-\tau^2_{\mathcal L}(\theta^*)δL(Δ,θ∗)≥κL​∥Δ∥2−τL2​(θ∗) for every Δ∈C(M,M‾⊥;θ∗)\Delta\in\mathbb C(\mathcal M,\overline{\mathcal M}^\perp;\theta^*)Δ∈C(M,M⊥;θ∗).

The conditions of the paper's main theorem are (G1): R\mathcal RR is a norm, decomposable with respect to (M,M‾⊥)(\mathcal M,\overline{\mathcal M}^\perp)(M,M⊥) with M⊆M‾\mathcal M\subseteq\overline{\mathcal M}M⊆M; and (G2): L\mathcal LL is convex, differentiable and satisfies RSC. The Lean development lives in the namespace UnifiedMEstimator.General, with these objects named IsNormFn, IsDecomposable, dualNorm, compat, setC, taylorErr, RSC and IsOptimal.

Formalization targets

Goal: Theorem 1 (p. 10), tolerance term corrected

Under (G1) and (G2), if λn>0\lambda_n>0λn​>0 and λn≥2R∗(∇L(θ∗))\lambda_n\ge2\mathcal R^*(\nabla\mathcal L(\theta^*))λn​≥2R∗(∇L(θ∗)), then every optimal solution of program (1) satisfies

∥θ^λn−θ∗∥2≤9 λn2κL2 Ψ2(M‾)+2τL2(θ∗)+4λnR(θM⊥∗)κL.\|\hat\theta_{\lambda_n}-\theta^*\|^2\le9\,\frac{\lambda_n^2}{\kappa_{\mathcal L}^2}\,\Psi^2(\overline{\mathcal M})+\frac{2\tau_{\mathcal L}^2(\theta^*)+4\lambda_n\mathcal R(\theta^*_{\mathcal M^\perp})}{\kappa_{\mathcal L}}.∥θ^λn​​−θ∗∥2≤9κL2​λn2​​Ψ2(M)+κL​2τL2​(θ∗)+4λn​R(θM⊥∗​)​.

The bound holds for every pair (M,M‾)(\mathcal M,\overline{\mathcal M})(M,M) over which R\mathcal RR decomposes, and for every optimum, not only a distinguished one.

Milestones

  1. Lemma 1 (p. 7): under the dual-norm condition on λn\lambda_nλn​, the error Δ^=θ^λn−θ∗\hat\Delta=\hat\theta_{\lambda_n}-\theta^*Δ^=θ^λn​​−θ∗ lies in C(M,M‾⊥;θ∗)\mathbb C(\mathcal M,\overline{\mathcal M}^\perp;\theta^*)C(M,M⊥;θ∗). This milestone links an existing platform statement of the same lemma (Wainwright, Proposition 9.13).
  2. Section 2.4, p. 10, first display: if θ∗∈M\theta^*\in\mathcal Mθ∗∈M and Δ∈C\Delta\in\mathbb CΔ∈C, then R(Δ)≤4Ψ(M‾)∥Δ∥\mathcal R(\Delta)\le4\Psi(\overline{\mathcal M})\|\Delta\|R(Δ)≤4Ψ(M)∥Δ∥.

Further statements

  • Corollary 1 (p. 11): if θ∗∈M\theta^*\in\mathcal Mθ∗∈M and τL(θ∗)=0\tau_{\mathcal L}(\theta^*)=0τL​(θ∗)=0, then ∥θ^λn−θ∗∥≤3λnΨ(M‾)/κL\|\hat\theta_{\lambda_n}-\theta^*\|\le3\lambda_n\Psi(\overline{\mathcal M})/\kappa_{\mathcal L}∥θ^λn​​−θ∗∥≤3λn​Ψ(M)/κL​ and R(θ^λn−θ∗)≤12λnΨ2(M‾)/κL\mathcal R(\hat\theta_{\lambda_n}-\theta^*)\le12\lambda_n\Psi^2(\overline{\mathcal M})/\kappa_{\mathcal L}R(θ^λn​​−θ∗)≤12λn​Ψ2(M)/κL​.
  • Section 2.4, p. 10, second display: a lower bound δL≥κ1∥Δ∥2−κ2g R2(Δ)\delta\mathcal L\ge\kappa_1\|\Delta\|^2-\kappa_2 g\,\mathcal R^2(\Delta)δL≥κ1​∥Δ∥2−κ2​gR2(Δ) on the unit ball gives curvature κ1−16κ2Ψ2(M‾)g\kappa_1-16\kappa_2\Psi^2(\overline{\mathcal M})gκ1​−16κ2​Ψ2(M)g on C\mathbb CC when θ∗∈M\theta^*\in\mathcal Mθ∗∈M.
  • Example 1 (p. 5) and the value Ψ(M(S))=∣S∣\Psi(\mathcal M(S))=\sqrt{|S|}Ψ(M(S))=∣S∣​ (p. 9): the ℓ1\ell_1ℓ1​-norm instance, which shows that the hypotheses of the goal can be met.

Significance

Theorem 1 reduces a consistency proof for a new regularized estimator to two checks: that the regularizer decomposes over a pair of subspaces adapted to the model, and that the loss is curved on the set C\mathbb CC, together with a bound on R∗(∇L(θ∗))\mathcal R^*(\nabla\mathcal L(\theta^*))R∗(∇L(θ∗)) that is usually a concentration inequality. The paper derives from it the slog⁡p/ns\log p/nslogp/n Lasso rate under restricted eigenvalue conditions, rates for weakly sparse (ℓq\ell_qℓq​-ball) vectors, and group-Lasso rates; companion papers use it for low-rank matrix estimation, matrix completion and generalized linear models. Because the bound holds for every pair (M,M‾)(\mathcal M,\overline{\mathcal M})(M,M), it gives an explicit trade-off between an estimation error and an approximation error R(θM⊥∗)\mathcal R(\theta^*_{\mathcal M^\perp})R(θM⊥∗​).

The theorem is proved in the paper's supplementary appendix. No machine-checked proof of it is known. On Prove2Me, Wainwright's textbook restatement (Theorem 9.19, HighDimStat.Decomposability.thm9_19_general_bound) is a related but different statement: its RSC condition is local, on a ball, with a tolerance proportional to R2(Δ)\mathcal R^2(\Delta)R2(Δ), and it has extra side conditions and a different bound. A formal proof of the present goal certifies the deterministic core that every corollary of the paper relies on.

Difficulty

The obvious argument compares the objective at θ^\hat\thetaθ^ and at θ∗\theta^*θ∗ and applies RSC to the error. RSC, however, is available only on the set C\mathbb CC, not on all of EEE: in high dimensions the loss is flat in many directions, so strong convexity fails. The work is to show first that the error lies in C\mathbb CC (Lemma 1, which rests on decomposability and the choice of λn\lambda_nλn​), and then to relate the regularizer to the error norm through the projections onto M‾\overline{\mathcal M}M and M‾⊥\overline{\mathcal M}^\perpM⊥. The distinction between M\mathcal MM and M‾\overline{\mathcal M}M matters throughout: the compatibility constant is taken on the larger space M‾\overline{\mathcal M}M, while the approximation error projects θ∗\theta^*θ∗ onto the complement of the smaller one. The bound comes from a quadratic inequality in ∥Δ^∥\|\hat\Delta\|∥Δ^∥, and the constants depend on how its terms are split.

Formalization scope

Representation. The parameter space is an arbitrary finite-dimensional real inner product space E (equivalently Rp\mathbb R^pRp with any inner product, as the paper allows); matrices are covered by the same abstraction. Subspaces are Submodule ℝ E, projections are Submodule.starProjection, and the gradient is Mathlib's gradient, under the hypothesis that L\mathcal LL is differentiable. The dual norm and Ψ\PsiΨ are real suprema (sSup). They equal the paper's quantities because R\mathcal RR is required to be a genuine norm (nonnegative, definite, absolutely homogeneous, subadditive) and EEE is finite-dimensional; Ψ({0})=0\Psi(\{0\})=0Ψ({0})=0. The tolerance is a real number τ\tauτ entering as τ2\tau^2τ2; RSC contains κ>0\kappa>0κ>0 and is quantified over exactly C(M,M‾⊥;θ∗)\mathbb C(\mathcal M,\overline{\mathcal M}^\perp;\theta^*)C(M,M⊥;θ∗) for the same pair and point as the decomposability. Every statement is for every optimal solution of program (1). The data Z1nZ_1^nZ1n​ are fixed and absorbed into L\mathcal LL, and θ∗\theta^*θ∗ is an arbitrary point: the paper's requirement that θ∗\theta^*θ∗ minimise the population risk is never used by the theorem and is dropped.

Corrections of the printed statements.

  1. Display (22) prints the tolerance term as λnκL⋅2τL2(θ∗)\frac{\lambda_n}{\kappa_{\mathcal L}}\cdot2\tau^2_{\mathcal L}(\theta^*)κL​λn​​⋅2τL2​(θ∗). As printed the statement is false: for E=RE=\mathbb RE=R, R=∣⋅∣\mathcal R=|\cdot|R=∣⋅∣, M=M‾=R\mathcal M=\overline{\mathcal M}=\mathbb RM=M=R, L(θ)=(max⁡(0,∣θ∣−1))2\mathcal L(\theta)=(\max(0,|\theta|-1))^2L(θ)=(max(0,∣θ∣−1))2, θ∗=0.9\theta^*=0.9θ∗=0.9, λn=0.01\lambda_n=0.01λn​=0.01, κL=1/2\kappa_{\mathcal L}=1/2κL​=1/2, τL2=10\tau^2_{\mathcal L}=10τL2​=10, the optimum is 000 and ∥Δ^∥2=0.81\|\hat\Delta\|^2=0.81∥Δ^∥2=0.81 exceeds the printed bound 0.40360.40360.4036. The goal states 2τL2(θ∗)/κL2\tau^2_{\mathcal L}(\theta^*)/\kappa_{\mathcal L}2τL2​(θ∗)/κL​; the two forms agree when τL=0\tau_{\mathcal L}=0τL​=0, and the constants 999 and 444 are the paper's.
  2. Corollary 1's (25a) prints ∥θ^−θ∗∥≤9λn2Ψ2(M‾)/κL\|\hat\theta-\theta^*\|\le9\lambda_n^2\Psi^2(\overline{\mathcal M})/\kappa_{\mathcal L}∥θ^−θ∗∥≤9λn2​Ψ2(M)/κL​, which fails for L(θ)=(θ−0.002)2\mathcal L(\theta)=(\theta-0.002)^2L(θ)=(θ−0.002)2, θ∗=0.001\theta^*=0.001θ∗=0.001, λn=0.004\lambda_n=0.004λn​=0.004 on R\mathbb RR; the mission states 3λnΨ(M‾)/κL3\lambda_n\Psi(\overline{\mathcal M})/\kappa_{\mathcal L}3λn​Ψ(M)/κL​. Its "C(M,M‾,θ∗)\mathbb C(\mathcal M,\overline{\mathcal M},\theta^*)C(M,M,θ∗)" is read as C(M,M‾⊥;θ∗)\mathbb C(\mathcal M,\overline{\mathcal M}^\perp;\theta^*)C(M,M⊥;θ∗).

Trivializations ruled out. A regularizer predicate weaker than a norm would let the real suprema collapse to the junk value 000 and make the λn\lambda_nλn​ condition or the Ψ\PsiΨ term free; RSC over all of EEE would be classical strong convexity, and RSC over the cone without the 4R(θM⊥∗)4\mathcal R(\theta^*_{\mathcal M^\perp})4R(θM⊥∗​) slack would make the goal false; decomposability without M⊆M‾\mathcal M\subseteq\overline{\mathcal M}M⊆M or with the bars misplaced changes the theorem. None of these is used. The ℓ1\ell_1ℓ1​ example and a checked one-dimensional instance show that all hypotheses of the goal can hold simultaneously.

Infrastructure and contributions. A complete development needs: Hölder's inequality for a norm and its dual norm, boundedness of the two suprema in finite dimension, the decomposability inequality R(θ∗+Δ)−R(θ∗)≥R(ΔM‾⊥)−R(ΔM‾)−2R(θM⊥∗)\mathcal R(\theta^*+\Delta)-\mathcal R(\theta^*)\ge\mathcal R(\Delta_{\overline{\mathcal M}^\perp})-\mathcal R(\Delta_{\overline{\mathcal M}})-2\mathcal R(\theta^*_{\mathcal M^\perp})R(θ∗+Δ)−R(θ∗)≥R(ΔM⊥​)−R(ΔM​)−2R(θM⊥∗​), the first-order characterization of convexity, and the solution of a scalar quadratic inequality. The dual-norm and compatibility-constant lemmas are reusable for every decomposable-regularizer mission. Proofs of the milestones, of the goal, and of the ℓ1\ell_1ℓ1​ instance are all welcome.

Selected references

  • S. N. Negahban, P. Ravikumar, M. J. Wainwright, B. Yu, A Unified Framework for High-Dimensional Analysis of M-Estimators with Decomposable Regularizers, Statistical Science 27(4), 2012, 538–557. arXiv:1010.2731v3, doi:10.1214/12-STS400
  • P. J. Bickel, Y. Ritov, A. B. Tsybakov, Simultaneous Analysis of Lasso and Dantzig Selector, Annals of Statistics 37(4), 2009, 1705–1732. arXiv:0801.1095
  • M. J. Wainwright, High-Dimensional Statistics: A Non-Asymptotic Viewpoint, Cambridge University Press, 2019, Chapter 9. doi:10.1017/9781108627771
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ProbabilityStatistics·Captain: mikedeng1

Adversarially Robust Generalization Requires More Data 1: A Robust-Error Lower Bound in the Gaussian ModelResearch Paper

Motivation

Classifiers trained to high standard accuracy on image benchmarks can be made to fail by perturbations of the input that are small in the ℓ∞\ell_\inftyℓ∞​ norm (Szegedy et al., 2014; Goodfellow et al., 2015). Adversarial training reaches high robust accuracy on the training set, but on CIFAR10 the robust accuracy on held-out data is much lower than on the training set (Madry et al., 2018). That is, robust generalization fails.

Schmidt, Santurkar, Tsipras, Talwar and Mądry (arXiv:1804.11285) ask whether this is a statistical phenomenon: does learning a robust classifier need more samples than learning an accurate one, even in the simplest distributional model? This mission formalizes their answer for a mixture of two Gaussians. In that model, with ∥θ⋆∥2=d\|\theta^\star\|_2 = \sqrt d∥θ⋆∥2​=d​ and σ≤c d1/4\sigma \le c\, d^{1/4}σ≤cd1/4, a single sample suffices for standard generalization (their Theorem 4). Robust generalization, by contrast, needs a number of samples that grows polynomially with the dimension, for every learning algorithm.

Setting

Write Rd\mathbb R^dRd for the feature space and {±1}\{\pm 1\}{±1} for the labels.

  • The ℓ∞\ell_\inftyℓ∞​ perturbation set of radius ε\varepsilonε around xxx is B∞ε(x)={x′∈Rd:∥x′−x∥∞≤ε}\mathcal B_\infty^\varepsilon(x) = \{x' \in \mathbb R^d : \|x' - x\|_\infty \le \varepsilon\}B∞ε​(x)={x′∈Rd:∥x′−x∥∞​≤ε}.
  • For θ∈Rd\theta \in \mathbb R^dθ∈Rd and σ>0\sigma > 0σ>0, the (θ,σ)(\theta, \sigma)(θ,σ)-Gaussian model Pθ,σP_{\theta,\sigma}Pθ,σ​ is the law of (x,y)(x, y)(x,y) obtained by drawing yyy uniformly from {±1}\{\pm1\}{±1} and then x∼N(yθ,σ2I)x \sim \mathcal N(y\theta, \sigma^2 I)x∼N(yθ,σ2I) (Definition 1). Here σ\sigmaσ is a standard deviation.
  • The ℓ∞ε\ell_\infty^\varepsilonℓ∞ε​-robust classification error of a classifier f:Rd→{±1}f : \mathbb R^d \to \{\pm1\}f:Rd→{±1} under a distribution PPP is P(x,y)∼P[∃ x′∈B∞ε(x):f(x′)≠y]\mathbb P_{(x,y) \sim P}[\exists\, x' \in \mathcal B_\infty^\varepsilon(x) : f(x') \ne y]P(x,y)∼P​[∃x′∈B∞ε​(x):f(x′)=y] (Definitions 2–3). With ε=0\varepsilon = 0ε=0 it is the ordinary classification error.
  • A learning algorithm gng_ngn​ maps nnn labelled samples S∈(Rd×{±1})nS \in (\mathbb R^d \times \{\pm 1\})^nS∈(Rd×{±1})n to a classifier fn=gn(S)f_n = g_n(S)fn​=gn​(S).
  • The expected robust error Ξ\XiΞ of gng_ngn​ is the robust error of gn(S)g_n(S)gn​(S) under Pθ,σP_{\theta,\sigma}Pθ,σ​, averaged over S∼Pθ,σ⊗nS \sim P_{\theta,\sigma}^{\otimes n}S∼Pθ,σ⊗n​ and then over a prior θ∼N(0,I)\theta \sim \mathcal N(0, I)θ∼N(0,I). The learner sees SSS but not θ\thetaθ.

Formalization targets

Goal: Corollary 23 (p. 30)

For every learning algorithm gng_ngn​, every σ>0\sigma > 0σ>0 and every ε≥0\varepsilon \ge 0ε≥0,

n≤ε2σ28log⁡d⟹Ξ ≥ (1−1d)12.n \le \frac{\varepsilon^2\sigma^2}{8\log d} \quad\Longrightarrow\quad \Xi \ \ge\ \Big(1 - \frac1d\Big)\frac12 .n≤8logdε2σ2​⟹Ξ ≥ (1−d1​)21​.

Theorem 11 (p. 28)

For every learning algorithm gng_ngn​, every σ>0\sigma > 0σ>0 and every ε≥0\varepsilon \ge 0ε≥0,

Ξ ≥ 12 Pv∼N(0,I)[nσ2+n ∥v∥∞≤ε].\Xi \ \ge\ \frac12\, \mathbb P_{v \sim \mathcal N(0, I)}\Big[\sqrt{\tfrac{n}{\sigma^2+n}}\,\|v\|_\infty \le \varepsilon\Big].Ξ ≥ 21​Pv∼N(0,I)​[σ2+nn​​∥v∥∞​≤ε].

Intermediate statements (milestones)

  1. Eq. (2). Given nnn samples zi∼N(θ,σ2I)z_i \sim \mathcal N(\theta, \sigma^2 I)zi​∼N(θ,σ2I), the posterior of θ∼N(0,I)\theta \sim \mathcal N(0, I)θ∼N(0,I) is N(μ′,Σ′)\mathcal N(\mu', \Sigma')N(μ′,Σ′) with μ′=(σ2+n)−1∑izi\mu' = (\sigma^2+n)^{-1}\sum_i z_iμ′=(σ2+n)−1∑i​zi​ and Σ′=σ2σ2+nI\Sigma' = \frac{\sigma^2}{\sigma^2+n} IΣ′=σ2+nσ2​I. The expectations over θ\thetaθ and over the samples may therefore be exchanged.
  2. Eq. (3). Averaging Pθ,σP_{\theta,\sigma}Pθ,σ​ over θ∼N(m,s2I)\theta \sim \mathcal N(m, s^2 I)θ∼N(m,s2I) gives Pm,s2+σ2P_{m, \sqrt{s^2+\sigma^2}}Pm,s2+σ2​​.
  3. The bound on Ψ\PsiΨ. If ∥m∥∞≤ε\|m\|_\infty \le \varepsilon∥m∥∞​≤ε, every classifier has ℓ∞ε\ell_\infty^\varepsilonℓ∞ε​-robust error at least 12\frac1221​ under Pm,sP_{m,s}Pm,s​.
  4. The law of zˉ\bar zzˉ. The sample mean zˉ\bar zzˉ of the ziz_izi​ is marginally N(0,(1+σ2/n)I)\mathcal N(0, (1+\sigma^2/n) I)N(0,(1+σ2/n)I).
  5. Maximum of ddd Gaussians. Pv∼N(0,Id)[∥v∥∞≤22log⁡d]≥1−1/d\mathbb P_{v\sim\mathcal N(0,I_d)}[\|v\|_\infty \le 2\sqrt{2\log d}] \ge 1 - 1/dPv∼N(0,Id​)​[∥v∥∞​≤22logd​]≥1−1/d.

Corollary 23 is the goal because it is the statement the paper advertises: its main-text Theorem 6 is Corollary 23 with σ=c1d1/4\sigma = c_1 d^{1/4}σ=c1​d1/4.

Significance

In the same model with ∥θ⋆∥2=d\|\theta^\star\|_2 = \sqrt d∥θ⋆∥2​=d​ and σ\sigmaσ of order d1/4d^{1/4}d1/4, a single sample suffices to reach standard error below 1% (Theorem 4 of the paper), and on the order of ε2d\varepsilon^2\sqrt dε2d​ samples suffice for robust error below 1% when ε\varepsilonε is below a small constant (Theorem 5; Corollary 22, formalized in mission 3 of this series). Corollary 23 shows that, up to the logarithmic factor, no learner can do better. Robust generalization then needs ε2d/log⁡d\varepsilon^2 \sqrt d / \log dε2d​/logd times as many samples as standard generalization. The gap is information-theoretic: it concerns every algorithm, not a particular training procedure or model class. The authors present this as a candidate explanation for the robust-generalization gap observed on CIFAR10. The ½ is tight: a constant classifier attains it.

The paper's proof is complete and short. As far as a search of the platform shows, none of its statements has been formalized. A machine-checked version requires multivariate Gaussian conjugacy, Gaussian convolution identities, the outer-measure robust event, and a union bound for the maximum of Gaussians. The Gaussian conjugacy and convolution facts are standard and appear throughout Bayesian statistics. As of this Mathlib version they are not available for stdGaussian on EuclideanSpace.

Difficulty

The obvious attempt fixes θ\thetaθ and bounds the robust error for each θ\thetaθ. That fails: a learner may ignore the data and output the Bayes-optimal robust classifier for one fixed θ\thetaθ, so for each θ\thetaθ some learner does well. The lower bound holds only on average over the prior on θ\thetaθ. The classifier fnf_nfn​ depends on the samples, and the samples depend on θ\thetaθ, so the classifier and the test distribution are correlated through θ\thetaθ. A second obstacle is measure-theoretic. The robust error of a classifier is the probability of an ℓ∞\ell_\inftyℓ∞​-thickening of an arbitrary set {f≠y}\{f \ne y\}{f=y}. Such a set need not be Borel, and it must be bounded below with no structure on the classifier beyond what the learner provides. The same statement with the ℓ2\ell_2ℓ2​ ball is a different theorem.

Formalization scope

  • Rd\mathbb R^dRd is EuclideanSpace ℝ (Fin d) with its Borel σ-algebra; N(0,I)\mathcal N(0, I)N(0,I) is Mathlib's stdGaussian; N(m,s2I)\mathcal N(m, s^2 I)N(m,s2I) is its image under v↦m+svv \mapsto m + s vv↦m+sv, so every Gaussian parameter in the development is a standard deviation.
  • The ℓ∞\ell_\inftyℓ∞​ ball is written coordinatewise (∣xi′−xi∣≤ε|x'_i - x_i| \le \varepsilon∣xi′​−xi​∣≤ε for all iii), because the ambient norm is ℓ2\ell_2ℓ2​. ∥v∥∞≤r\|v\|_\infty \le r∥v∥∞​≤r is written the same way.
  • Labels are Bool, with true for +1+1+1. A classifier is ℝ^d → Bool, and a learning algorithm is (Fin n → ℝ^d × Bool) → ℝ^d → Bool.
  • A model is a measure on Rd×\mathbb R^d \timesRd× Bool; nnn samples form the product measure Measure.pi.
  • The robust event need not be Borel. Its probability is the outer measure, which is its probability under the completion. The expectations are lower Lebesgue integrals.
  • Theorem 11 and Corollary 23 assume the learner is jointly measurable in (samples, input). This is the only condition on it. log⁡\loglog is the natural logarithm. With Lean's conventions log⁡0=log⁡1=0\log 0 = \log 1 = 0log0=log1=0 and x/0=0x/0 = 0x/0=0, the goal's hypothesis forces n=0n = 0n=0 for d≤1d \le 1d≤1, where the statement is still true.
  • The paper writes the posterior mean as nσ2+nzˉ\frac{n}{\sigma^2+n}\bar zσ2+nn​zˉ, with "zˉ=∑izi\bar z = \sum_i z_izˉ=∑i​zi​" on p. 28. The formalization uses (σ2+n)−1∑izi(\sigma^2+n)^{-1}\sum_i z_i(σ2+n)−1∑i​zi​, which is the posterior mean. It agrees with the paper when zˉ\bar zzˉ is read as the sample mean, as it is on p. 30.

A formalization that fixes θ\thetaθ instead of averaging over the prior, that restricts the learner (to linear classifiers, or to classifiers that do not depend on the data), that uses the ℓ2\ell_2ℓ2​ ball, or that assumes the posterior formula as a hypothesis states a different theorem. These are ruled out by the definitions file.

Needed infrastructure: the multivariate Gaussian conjugacy and convolution identities for stdGaussian pushforwards, translation invariance of outer measure under Gaussian shifts, and a sub-Gaussian tail bound for one coordinate. The Gaussian identities are reusable well beyond this mission. Contributions of general Gaussian lemmas, stated for stdGaussian on any finite-dimensional inner product space, are welcome. Related platform work: missions 2–4 of this series formalize the Bernoulli-model lower bound and the two upper bounds of the same paper.

Selected references

  • L. Schmidt, S. Santurkar, D. Tsipras, K. Talwar, A. Mądry, Adversarially Robust Generalization Requires More Data, NeurIPS 2018; arXiv:1804.11285v2, 2018. https://arxiv.org/abs/1804.11285
  • A. Madry, A. Makelov, L. Schmidt, D. Tsipras, A. Vladu, Towards Deep Learning Models Resistant to Adversarial Attacks, ICLR 2018. https://arxiv.org/abs/1706.06083
  • C. Szegedy, W. Zaremba, I. Sutskever, J. Bruna, D. Erhan, I. Goodfellow, R. Fergus, Intriguing properties of neural networks, ICLR 2014. https://arxiv.org/abs/1312.6199
  • I. Goodfellow, J. Shlens, C. Szegedy, Explaining and Harnessing Adversarial Examples, ICLR 2015. https://arxiv.org/abs/1412.6572
  • S. Boucheron, G. Lugosi, P. Massart, Concentration Inequalities: A Nonasymptotic Theory of Independence, Oxford University Press, 2013 (Theorem 5.8). https://doi.org/10.1093/acprof:oso/9780199535255.001.0001
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Algorithmic Game TheoryProbabilityStatistics·Captain: mikedeng1

Calibrated Learning and Correlated Equilibrium III: A Randomized Forecast Calibrated against Every OpponentResearch Paper

Motivation

A forecaster who announces "70% chance of rain" is calibrated if, among the days on which that number was announced, it rained on about 70% of them. Dawid (The well-calibrated Bayesian, JASA 1982) proposed calibration as the minimal requirement of an honest probability forecaster. Oakes (Self-calibrating priors do not exist, JASA 1985) showed that no deterministic forecasting rule can be calibrated against every sequence of outcomes: an adversary who knows the rule can always choose the outcome that contradicts the forecast.

Foster and Vohra (Calibrated learning and correlated equilibrium, Games Econ. Behav. 21 (1997) 40–55) use calibration as the bridge between learning and equilibrium in repeated games. Their Theorem 1 says that if each player best-responds to calibrated forecasts of the opponent, the empirical distribution of play converges to the set of correlated equilibria. That theorem is only useful if calibrated forecasts can actually be produced whatever the opponent does. Theorem 3 of the paper, credited to an unpublished 1991 manuscript of the same authors and proved in the paper's Appendix, says they can, provided the forecaster randomizes.

Timeline. Dawid (1982) defines calibration. Oakes (1985) rules out deterministic calibrated forecasting against arbitrary sequences. Foster and Vohra (1991 manuscript; 1997 paper, Theorem 3 and Appendix) give a randomized forecaster calibrated against any opponent, through a pairwise ("internal") no-regret property. The full argument appeared in Foster and Vohra, Asymptotic calibration, Biometrika 85 (1998). Hart and Mas-Colell (A simple adaptive procedure leading to correlated equilibrium, Econometrica 2000) later made internal regret the standard route to correlated equilibrium.

Setting

Player 2 has n≥1n\ge 1n≥1 pure strategies j∈{0,…,n−1}j\in\{0,\dots,n-1\}j∈{0,…,n−1}. In every round, player 1 announces a forecast p∈Rnp\in\mathbb R^np∈Rn, a probability vector (pj≥0p_j\ge 0pj​≥0, ∑jpj=1\sum_j p_j = 1∑j​pj​=1), and player 2 plays a strategy jjj. The two moves are simultaneous: player 2 does not see the current forecast.

A history hhh of length ttt is the list of the ttt pairs (forecast, play), oldest first. For a forecast vector ppp and a strategy jjj:

  • N(p,t)N(p,t)N(p,t) is the number of rounds of hhh in which ppp was forecast;
  • ρ(p,j,t)\rho(p,j,t)ρ(p,j,t) is the fraction of those rounds in which player 2 played jjj (and 000 if N(p,t)=0N(p,t)=0N(p,t)=0);
  • the calibration score (Eq. (1), p. 49) is
Ct=∑p∑j∣ρ(p,j,t)−pj∣ N(p,t)t.C_t = \sum_p\sum_j \bigl|\rho(p,j,t) - p_j\bigr|\,\frac{N(p,t)}{t}.Ct​=p∑​j∑​​ρ(p,j,t)−pj​​tN(p,t)​.

A randomized forecaster FFF maps each history to a probability distribution on forecasts. A learning rule AAA of player 2 maps each history to a probability distribution on strategies. In round t+1t+1t+1 the forecast is drawn from F(h)F(h)F(h) and the play from A(h)A(h)A(h), independently given the history hhh of the first ttt rounds. This defines the law PF,A\mathbb P_{F,A}PF,A​ of the first ttt rounds (histLaw F A t).

For the Appendix: with kkk forecasts, losses LtiL_t^iLti​ and mixing weights wtiw_t^iwti​, the pairwise regret of replacing forecast iii by forecast jjj is

RTi→j=max⁡{0, ∑t=1Twti (Lti−Ltj)}.R_T^{i\to j} = \max\Bigl\{0,\ \sum_{t=1}^T w_t^i\,(L_t^i - L_t^j)\Bigr\}.RTi→j​=max{0, t=1∑T​wti​(Lti​−Ltj​)}.

Formalization targets

Goal: Theorem 3 (p. 49)

There is a forecaster FFF, with probability-vector forecasts, such that for every learning rule AAA of player 2 and every ε>0\varepsilon>0ε>0,

lim⁡t→∞PF,A(Ct<ε)=1.\lim_{t\to\infty}\mathbb P_{F,A}\bigl(C_t<\varepsilon\bigr) = 1.t→∞lim​PF,A​(Ct​<ε)=1.

The forecaster is fixed before the opponent; no rate is claimed, and the rate may depend on AAA.

Milestones, in the order the Appendix uses them

  1. Flow conservation is solvable (p. 52). For every nonnegative k×kk\times kk×k matrix RRR, k≥1k\ge 1k≥1, there is a probability vector www with wi∑jRi→j=∑jwjRj→iw^i\sum_j R^{i\to j} = \sum_j w^j R^{j\to i}wi∑j​Ri→j=∑j​wjRj→i for all iii.
  2. Lemma 1 (No-Regret) (p. 52). With losses in [0,1][0,1][0,1] and weights solving flow conservation for the previous regrets,
RTi→j≤2kTfor all i,j,T.R_T^{i\to j}\le\sqrt{2kT}\quad\text{for all } i,j,T.RTi→j​≤2kT​for all i,j,T.
  1. Regrets sandwich L-2 calibration (p. 54). For a grid p1,…,pkp^1,\dots,p^kp1,…,pk that is ε\varepsilonε-dense in squared distance and losses Lti=∣Xt−pi∣2L_t^i = |X_t - p^i|^2Lti​=∣Xt​−pi∣2,
∑imax⁡jRTi→jT ≤ C2,w(T) ≤ ε+∑imax⁡jRTi→jT,\sum_i\max_j \frac{R_T^{i\to j}}{T}\ \le\ C_{2,w}(T)\ \le\ \varepsilon + \sum_i\max_j\frac{R_T^{i\to j}}{T},i∑​jmax​TRTi→j​​ ≤ C2,w​(T) ≤ ε+i∑​jmax​TRTi→j​​,

with C2,wC_{2,w}C2,w​ the fractional L-2 calibration score. 4. L-1 versus L-2 (p. 54). For each jjj, ∑p∣ρ(p,j,t)−pj∣ N(p,t)/t≤∑p(ρ(p,j,t)−pj)2N(p,t)/t\sum_p|\rho(p,j,t)-p_j|\,N(p,t)/t \le \sqrt{\sum_p(\rho(p,j,t)-p_j)^2 N(p,t)/t}∑p​∣ρ(p,j,t)−pj​∣N(p,t)/t≤∑p​(ρ(p,j,t)−pj​)2N(p,t)/t​.

Significance

The result. Theorem 3 makes the hypothesis of Theorem 1 attainable: combined, they show that there are learning procedures under which play converges in probability to the set of correlated equilibria of any finite game (the paper's Corollary, p. 49). The intermediate Lemma 1 is an early internal-regret bound; internal (swap) regret minimization later became the standard algorithmic route to correlated equilibria and to calibrated prediction in online learning.

Formalizing it. The result is proved, in the 1997 Appendix in telegraphic form and in full in Foster and Vohra (1998). The Appendix leaves several steps informal (see Formalization scope), so a machine-checked proof must supply them. No formalization of calibration or of internal regret was found on Prove2Me as of 2026-09-26. The mission produces a formal model of randomized forecasters against adaptive opponents, a checked internal-regret bound with an explicit constant, and the passage from regret to calibration.

Difficulty

The obvious approach is to pick, at each round, a forecast that corrects the current miscalibration. This is a deterministic rule, and by Oakes' theorem an opponent can defeat it. Randomization alone does not help either: the forecaster must randomize in a way that controls every pairwise regret Ri→jR^{i\to j}Ri→j at once, not only the regret against the best fixed forecast. External no-regret does not imply calibration.

Two further gaps separate Lemma 1 from Theorem 3. First, the Appendix controls a fractional score in which the event "forecast pip^ipi was issued" is replaced by its probability wtiw_t^iwti​. The realized calibration score involves the random choices, so a concentration argument is needed against an adaptive opponent. Second, a fixed grid gives calibration only up to its mesh ε\varepsilonε. Exact convergence requires letting the grid size kkk grow and ε\varepsilonε shrink over time, and the scores of the different phases must be combined.

Formalization scope

  • Model. Strategies are Fin n; forecasts are vectors Fin n → ℝ that are probability vectors (IsDist). Histories are Lean lists of (forecast, play) pairs, oldest first. Forecaster and opponent are maps from histories to Mathlib PMFs. The history law is built with PMF.bind/PMF.map, with the two draws independent given the past. P(Ct<ε)\mathbb P(C_t<\varepsilon)P(Ct​<ε) is the toOuterMeasure of the law, in ℝ≥0∞; no σ-algebra on histories is used.
  • Opponent. The opponent may be randomized and may depend on all past forecasts and plays; fixed sequences and deterministic rules are special cases. The opponent never sees the current forecast. Letting it see the current forecast would make the goal false, and restricting to fixed sequences would make it weaker than the paper.
  • Quantifiers. The forecaster is chosen before the opponent (∃ F, ∀ A). The reverse order is trivial, since one can forecast AAA's next play.
  • Conventions. Rounds are counted from 000, so the paper's rounds 1,…,t1,\dots,t1,…,t are the first ttt list entries. The paper's Rt−1R_{t-1}Rt−1​ is the regret over the 000-based rounds before ttt. Forecasts are grouped by exact equality of real vectors. Sums over ppp run over the forecasts that occur, and every other term vanishes. Scores divide by the history length, and are 000 for the empty history. "Converges in probability" is the lim⁡P(Ct<ε)=1\lim\mathbb P(C_t<\varepsilon)=1limP(Ct​<ε)=1 form the page states. The grid of milestone 3 is indexed 1,…,k1,\dots,k1,…,k (the page writes i=0,…,ki = 0,\dots,ki=0,…,k on p. 53 and 1,…,k1,\dots,k1,…,k in Lemma 1), and "within ε\varepsilonε" is read in squared Euclidean distance.
  • Pinned reading. The page writes the middle term of milestone 3 as E(C2(t))E(C_2(t))E(C2​(t)) with a garbled formula. The mission states the inequality for the fractional score C2,wC_{2,w}C2,w​, for which it holds.
  • Steps the paper leaves informal, not stated as milestones. (a) "E(C2(t))≤ε+O(k/2)E(C_2(t))\le\varepsilon + O(k/\sqrt2)E(C2​(t))≤ε+O(k/2​)" when the weights solve flow conservation; the OOO-term is garbled and should decay in ttt. (b) "if we let kkk grow slowly and ε\varepsilonε go slowly to zero … C2(t)→0C_2(t)\to 0C2​(t)→0 in expectation which implies C2(t)→0C_2(t)\to0C2​(t)→0 in probability by Jensen's inequality", together with the passage from the fractional score to the realized one. Solvers will have to formalize these steps on the way to the goal.
  • Not included. The Corollary on p. 49 (convergence in probability of play to the correlated equilibria when both players use the scheme). It needs the game layer and a quantitative form of Theorem 1, and the page gives no proof.
  • Infrastructure welcome. Finite Markov chain stationary distributions (for milestone 1), for which Prove2Me has MarkovChain.exists_isStationary for row-stochastic matrices. Also useful: martingale concentration for PMF-built processes, and Cauchy–Schwarz with weights. The history-law construction is reusable for any repeated forecasting game.

Selected references

  • D. P. Foster and R. V. Vohra, Calibrated learning and correlated equilibrium, Games and Economic Behavior 21 (1997) 40–55. https://doi.org/10.1006/game.1997.0595
  • D. P. Foster and R. V. Vohra, Asymptotic calibration, Biometrika 85 (1998) 379–390. https://doi.org/10.1093/biomet/85.2.379
  • A. P. Dawid, The well-calibrated Bayesian, J. Amer. Statist. Assoc. 77 (1982) 605–613. https://doi.org/10.1080/01621459.1982.10477856
  • D. Oakes, Self-calibrating priors do not exist, J. Amer. Statist. Assoc. 80 (1985) 339. https://doi.org/10.1080/01621459.1985.10478117
  • S. Hart and A. Mas-Colell, A simple adaptive procedure leading to correlated equilibrium, Econometrica 68 (2000) 1127–1150. https://doi.org/10.1111/1468-0262.00153
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Algorithmic Game TheoryOperations Research·Captain: mikedeng1

Calibrated Learning and Correlated Equilibrium I: Calibrated Forecasts with Best Responses Converge to the Set of Correlated EquilibriaResearch Paper

Motivation

A correlated equilibrium (Aumann 1974) is a joint distribution over the players' strategy profiles such that no player gains by deviating from the strategy the distribution recommends to them. It is the equilibrium notion that learning dynamics in repeated games most naturally reach, and a basic question in learning in games is which simple rules, played repeatedly, drive the empirical distribution of play to the set of correlated equilibria.

Foster and Vohra (1997) answer this with a hypothesis on forecasts instead of a particular algorithm. Each player forecasts the other's next move and best-responds to the forecast. They only require the forecasts to be calibrated in the sense of Dawid (1982): among the rounds in which a player forecast a given probability vector, the empirical frequencies of the opponent's moves must approach that vector. Their Theorem 1 says that this already forces the empirical joint distribution of play to approach the set of correlated equilibria. The paper uses this to argue that Bayesian players under a common prior, whose forecasts are calibrated by Dawid's theorem, end up playing a correlated equilibrium. That is an alternative to Aumann's (1987) derivation of correlated equilibrium from common priors and rationality.

Timeline:

  • Aumann (1974, 1987) introduces correlated equilibrium and derives it from Bayesian rationality.
  • Dawid (1982) proposes calibration as a minimal requirement on probability forecasts.
  • Foster and Vohra (1997) prove Theorem 1 (this mission) and show that calibrated forecasts exist once the forecaster may randomize.
  • Hart and Mas-Colell (2000) give regret matching, an adaptive procedure with the same limit set.

Setting

A finite two-player game GGG has strategy sets S(1)={0,…,m−1}S(1) = \{0, \dots, m-1\}S(1)={0,…,m−1} and S(2)={0,…,n−1}S(2) = \{0, \dots, n-1\}S(2)={0,…,n−1} and payoff matrices u1,u2:S(1)×S(2)→Ru_1, u_2 : S(1) \times S(2) \to \mathbb{R}u1​,u2​:S(1)×S(2)→R, which the players maximize. A joint distribution DDD is a nonnegative m×nm \times nm×n matrix with entries summing to 111. It is a correlated equilibrium if

∑x,yD(x,y) u1(Φ(x),y)≤∑x,yD(x,y) u1(x,y)for all Φ:S(1)→S(1),\sum_{x,y} D(x,y)\, u_1(\Phi(x), y) \le \sum_{x,y} D(x,y)\, u_1(x,y) \quad \text{for all } \Phi : S(1) \to S(1),x,y∑​D(x,y)u1​(Φ(x),y)≤x,y∑​D(x,y)u1​(x,y)for all Φ:S(1)→S(1),

and symmetrically for player 2. The set of correlated equilibria is π(G)\pi(G)π(G).

The game is played in rounds s=0,1,2,…s = 0, 1, 2, \dotss=0,1,2,…. In round sss player 1 issues a forecast f1(s)f_1(s)f1​(s), a probability vector over S(2)S(2)S(2), and player 2 issues a forecast f2(s)f_2(s)f2​(s) over S(1)S(1)S(1). Each player then plays a best response to its forecast, x(s)=R1(f1(s))x(s) = R_1(f_1(s))x(s)=R1​(f1​(s)) and y(s)=R2(f2(s))y(s) = R_2(f_2(s))y(s)=R2​(f2​(s)). Here R1R_1R1​ and R2R_2R2​ are best-reply functions: R1(p)R_1(p)R1​(p) maximizes ∑ypyu1(⋅,y)\sum_y p_y u_1(\cdot, y)∑y​py​u1​(⋅,y) for every probability vector ppp, and R1R_1R1​ is a fixed function of the forecast alone. This is the paper's standing assumption of a stationary, deterministic tie-breaking rule.

For a forecast sequence fff and the opponent's plays zzz, N(p,t)N(p,t)N(p,t) counts the rounds among the first ttt in which fff forecast ppp. ρ(p,j,t)\rho(p,j,t)ρ(p,j,t) is the fraction of those rounds in which the opponent played jjj, and 000 if there are none. The forecast is calibrated with respect to zzz if for every jjj

∑p∣ρ(p,j,t)−pj∣ N(p,t)t⟶0(t→∞).\sum_p |\rho(p,j,t) - p_j|\, \frac{N(p,t)}{t} \longrightarrow 0 \qquad (t \to \infty).p∑​∣ρ(p,j,t)−pj​∣tN(p,t)​⟶0(t→∞).

The empirical joint distribution Dt(x,y)D_t(x,y)Dt​(x,y) is the fraction of the first ttt rounds in which player 1 played xxx and player 2 played yyy.

Formalization targets

Goal: Theorem 1

If f1f_1f1​ is calibrated with respect to yyy and f2f_2f2​ is calibrated with respect to xxx, then

min⁡D∈π(G) max⁡x∈S(1), y∈S(2)∣Dt(x,y)−D(x,y)∣⟶0(t→∞).\min_{D \in \pi(G)} \ \max_{x \in S(1),\, y \in S(2)} |D_t(x,y) - D(x,y)| \longrightarrow 0 \qquad (t \to \infty).D∈π(G)min​ x∈S(1),y∈S(2)max​∣Dt​(x,y)−D(x,y)∣⟶0(t→∞).

The goal fixes no rate and no particular forecasting method: it asserts only convergence of DtD_tDt​ to the set π(G)\pi(G)π(G), for every calibrated forecast.

Milestones: the steps of the proof (pp. 44–45)

  1. DtD_tDt​ lies in the simplex for t≥1t \ge 1t≥1.
  2. For each x∈S(1)x \in S(1)x∈S(1), the set Mb(x)M_b(x)Mb​(x) of mixtures to which xxx is a best response is closed and convex.
  3. The mixtures Mp(x)M_p(x)Mp​(x) at which player 1 actually plays xxx satisfy Mp(x)⊆Mb(x)M_p(x) \subseteq M_b(x)Mp​(x)⊆Mb​(x).
  4. The identity writing Dt(x,y)D_t(x,y)Dt​(x,y) as a forecast-weighted term plus a calibration error.
  5. Calibration makes the error term vanish.
  6. The weighted average of the forecasts at which player 1 plays xxx lies in Mb(x)M_b(x)Mb​(x).
  7. For a convergent subsequence Dti→DD_{t_i} \to DDti​​→D, every row of DDD with positive mass, normalized, lies in Mb(x)M_b(x)Mb​(x).
  8. Every subsequential limit of DtD_tDt​ is a correlated equilibrium.

Further result: matching pennies (p. 46)

With the constant forecast (1/2,1/2)(1/2, 1/2)(1/2,1/2) and the non-stationary tie-break "heads on even rounds, tails on odd rounds", both forecasts are calibrated and every play is a best reply, yet DtD_tDt​ does not approach π(G)\pi(G)π(G). The stationarity assumption cannot be dropped.

Significance

The result. Theorem 1 separates what learning needs from how it is achieved. Any forecasting procedure that is calibrated, combined with myopic best responses, yields correlated equilibrium behaviour in the long run. The paper's Theorem 3 constructs a randomized calibrated forecaster, so the theorem gives an uncoupled learning procedure for correlated equilibrium, one that needs no knowledge of the opponent's payoffs. The converse direction, that every correlated equilibrium arises this way for almost every game, is the paper's Theorem 2 (a separate mission of this series).

Formalizing it. The theorem is proved in the paper and has no machine-checked proof on Prove2Me or, to the knowledge of this mission, elsewhere. The platform's existing correlated-equilibrium results (the Algorithmic Game Theory swap-regret development, AGT.swap_regret_correlated_equilibrium) reach correlated equilibrium through swap regret of mixed strategies, a different hypothesis and a different object. This mission adds a formal notion of calibration and the convergence argument, both reusable for the paper's Theorems 2 and 3 and for later work on calibration and learning.

Difficulty

The obvious reading of calibration is that each player's forecast converges to the opponent's empirical distribution; if that held, best responses to it would give convergence. It does not hold. Calibration constrains the opponent's frequencies only conditionally on the forecast issued, and the forecasts need not converge at all. What must be shown is a statement about the conditional distributions of the joint play given each strategy of player 1, while the set of forecasts issued keeps growing. Rows of the limit with zero mass carry no conditional distribution. The "min → 0" form also asks for more than a property of limit points: it is a uniform statement about all large ttt.

Formalization scope

  • Strategies are Fin m and Fin n; payoffs are real matrices; forecasts are real vectors required to be probability vectors in every round.
  • A correlated equilibrium is the joint-distribution form of p. 44 (the correlated strategy on a finite probability space is represented by its law). It is the ε=0\varepsilon = 0ε=0, two-player, payoff (not cost) instance of the published AGT.IsCorrelatedEquilibrium, restated rather than imported.
  • The stationary deterministic tie-break is modelled by arbitrary best-reply functions RiR_iRi​ of the forecast. They do not depend on the round, and the statements quantify over all of them, which includes the lowest-index rule.
  • Forecasts are sequences fi:N→Rkf_i : \mathbb{N} \to \mathbb{R}^kfi​:N→Rk. The theorem uses only the realized forecasts, and every sequence is realized by a rule reading the round number from the history.
  • Rounds are indexed from 000: "the first ttt rounds" are 0,…,t−10, \dots, t-10,…,t−1. D0=0D_0 = 0D0​=0 by Lean's division convention; only t≥1t \ge 1t≥1 and limits are used.
  • The calibration sum runs over the forecasts issued in the first ttt rounds, which is the paper's sum over all ppp with its zero terms removed. ρ(p,j,t)=0\rho(p,j,t) = 0ρ(p,j,t)=0 when N(p,t)=0N(p,t) = 0N(p,t)=0, as on the page.
  • "min … → 0" is stated as: for every ε>0\varepsilon > 0ε>0, eventually some D∈π(G)D \in \pi(G)D∈π(G) is within ε\varepsilonε of DtD_tDt​ in every coordinate. The two forms are equivalent because π(G)\pi(G)π(G) is compact and nonempty. The formalization avoids an infimum over π(G)\pi(G)π(G), which Lean would evaluate to 000 on an empty set.
  • A statement that drops the best-reply property, the probability-vector condition on forecasts, or the stationarity of RiR_iRi​ is not Theorem 1: the matching pennies example shows the last one is essential. Swapping the calibration hypotheses (player 1's forecast calibrated against player 1's own plays) type-checks when m=nm = nm=n and is not the theorem.
  • Only the two-player case is claimed. The paper says the results "generalize easily to the nnn-person case" without proof.

Proofs of any milestone, and reusable lemmas about calibration scores and compactness of the simplex of joint distributions, are welcome.

Selected references

  • D. P. Foster, R. V. Vohra, Calibrated learning and correlated equilibrium, Games and Economic Behavior 21 (1997) 40–55. https://doi.org/10.1006/game.1997.0595
  • R. J. Aumann, Subjectivity and correlation in randomized strategies, Journal of Mathematical Economics 1 (1974) 67–96. https://doi.org/10.1016/0304-4068(74)90037-8
  • R. J. Aumann, Correlated equilibrium as an expression of Bayesian rationality, Econometrica 55 (1987) 1–18. https://doi.org/10.2307/1911154
  • A. P. Dawid, The well-calibrated Bayesian, Journal of the American Statistical Association 77 (1982) 605–610. https://doi.org/10.1080/01621459.1982.10477856
  • S. Hart, A. Mas-Colell, A simple adaptive procedure leading to correlated equilibrium, Econometrica 68 (2000) 1127–1150. https://doi.org/10.1111/1468-0262.00153
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OptimizationProbabilityStatistics·Captain: mikedeng1

Simultaneous Analysis of Lasso and Dantzig Selector V: Estimation, Prediction and Sparsity Bounds for the LassoResearch Paper

Motivation

In a linear regression with many more candidate variables than observations, least squares is not defined uniquely and does not estimate anything useful. The Lasso (Tibshirani, 1996) replaces it by an ℓ1\ell_1ℓ1​-penalised least-squares problem, which is convex, can be solved at scale, and returns sparse coefficient vectors. The question that a statistician, a signal-processing engineer or an operations researcher fitting a sparse model must answer before trusting it is quantitative: how far is the Lasso estimate from the true coefficient vector, how well does it predict, and how many variables does it select, as functions of the sample size nnn, the number of variables MMM and the sparsity sss of the truth?

Bickel, Ritov and Tsybakov (arXiv:0801.1095, Ann. Statist. 37(4), 2009) answered this under the restricted eigenvalue (RE) condition, which they introduced, with explicit constants and an explicit failure probability. Their Theorem 7.2, the goal of this mission, is a standard reference result of high-dimensional statistics and a model for the Lasso analyses in the textbooks of Bühlmann and van de Geer (2011) and Wainwright (2019).

Timeline, restricted to what each work proved:

  • 2007: Candès and Tao (arXiv:math/0506081) prove ℓ2\ell_2ℓ2​ bounds for the Dantzig selector under a uniform uncertainty principle.
  • 2007: Bunea, Tsybakov and Wegkamp (doi:10.1214/07-EJS008) prove sparsity oracle inequalities for the Lasso under mutual-coherence conditions; Lemma B.1 of the present paper is essentially their Lemma 1.
  • 2008/2009: Bickel, Ritov and Tsybakov prove Theorem 7.2 under RE(s,3)(s,3)(s,3) and RE(s,m,3)(s,m,3)(s,m,3), conditions weaker than those of the previous works.

Setting

A deterministic design matrix X∈Rn×MX\in\mathbb R^{n\times M}X∈Rn×M with columns x(1),…,x(M)x_{(1)},\dots,x_{(M)}x(1)​,…,x(M)​ is observed together with

y=Xβ∗+w,y=X\beta^*+w,y=Xβ∗+w,

where β∗∈RM\beta^*\in\mathbb R^Mβ∗∈RM is unknown and w=(W1,…,Wn)w=(W_1,\dots,W_n)w=(W1​,…,Wn​) has independent N(0,σ2)\mathcal N(0,\sigma^2)N(0,σ2) entries with σ>0\sigma>0σ>0. Throughout, n≥1n\ge1n≥1, M≥2M\ge2M≥2, and every diagonal entry of the Gram matrix Ψn=X⊤X/n\Psi_n=X^\top X/nΨn​=X⊤X/n equals 111.

For δ∈RM\delta\in\mathbb R^Mδ∈RM write ∣δ∣p=(∑j∣δj∣p)1/p|\delta|_p=(\sum_j|\delta_j|^p)^{1/p}∣δ∣p​=(∑j​∣δj​∣p)1/p, J(δ)={j:δj≠0}J(\delta)=\{j:\delta_j\ne0\}J(δ)={j:δj​=0} for the support, M(δ)=∣J(δ)∣\mathcal M(\delta)=|J(\delta)|M(δ)=∣J(δ)∣ for the sparsity, and δJ\delta_JδJ​ for the vector that agrees with δ\deltaδ on JJJ and vanishes off JJJ. The largest eigenvalue of Ψn\Psi_nΨn​ is ϕmax⁡\phi_{\max}ϕmax​.

The Lasso estimator with tuning parameter r>0r>0r>0 is any minimiser

β^L∈arg⁡min⁡β∈RM{1n∣y−Xβ∣22+2r∣β∣1}.\hat\beta_L\in\arg\min_{\beta\in\mathbb R^M}\Big\{\frac1n|y-X\beta|_2^2+2r|\beta|_1\Big\}.β^​L​∈argβ∈RMmin​{n1​∣y−Xβ∣22​+2r∣β∣1​}.

Minimisers exist but need not be unique.

Assumption RE(s,c0)(s,c_0)(s,c0​) (1≤s≤M1\le s\le M1≤s≤M, c0>0c_0>0c0​>0) asks that

κ(s,c0)=min⁡∣J0∣≤s min⁡δ≠0, ∣δJ0c∣1≤c0∣δJ0∣1∣Xδ∣2n ∣δJ0∣2>0.\kappa(s,c_0)=\min_{|J_0|\le s}\ \min_{\delta\ne0,\ |\delta_{J_0^c}|_1\le c_0|\delta_{J_0}|_1}\frac{|X\delta|_2}{\sqrt n\,|\delta_{J_0}|_2}>0 .κ(s,c0​)=∣J0​∣≤smin​ δ=0, ∣δJ0c​​∣1​≤c0​∣δJ0​​∣1​min​n​∣δJ0​​∣2​∣Xδ∣2​​>0.

Assumption RE(s,m,c0)(s,m,c_0)(s,m,c0​) (1≤s≤M/21\le s\le M/21≤s≤M/2, m≥sm\ge sm≥s, s+m≤Ms+m\le Ms+m≤M) is the same with ∣δJ01∣2|\delta_{J_{01}}|_2∣δJ01​​∣2​ in the denominator, where J01=J0∪J1J_{01}=J_0\cup J_1J01​=J0​∪J1​ and J1J_1J1​ collects the mmm largest in absolute value coordinates of δ\deltaδ outside J0J_0J0​.

Formalization targets

Goal: Theorem 7.2

Let M(β∗)≤s\mathcal M(\beta^*)\le sM(β∗)≤s, let RE(s,3)(s,3)(s,3) hold, and let r=Aσlog⁡M/nr=A\sigma\sqrt{\log M/n}r=AσlogM/n​ with A>22A>2\sqrt2A>22​. With probability at least 1−M1−A2/81-M^{1-A^2/8}1−M1−A2/8, every Lasso solution satisfies

∣β^L−β∗∣1≤16Aκ2(s,3)σslog⁡Mn,∣X(β^L−β∗)∣22≤16A2κ2(s,3)σ2slog⁡M,M(β^L)≤64ϕmax⁡κ2(s,3)s,|\hat\beta_L-\beta^*|_1\le\frac{16A}{\kappa^2(s,3)}\sigma s\sqrt{\frac{\log M}{n}},\qquad |X(\hat\beta_L-\beta^*)|_2^2\le\frac{16A^2}{\kappa^2(s,3)}\sigma^2s\log M,\qquad \mathcal M(\hat\beta_L)\le\frac{64\phi_{\max}}{\kappa^2(s,3)}s,∣β^​L​−β∗∣1​≤κ2(s,3)16A​σsnlogM​​,∣X(β^​L​−β∗)∣22​≤κ2(s,3)16A2​σ2slogM,M(β^​L​)≤κ2(s,3)64ϕmax​​s,

and, if RE(s,m,3)(s,m,3)(s,m,3) holds, on the same event and for all 1<p≤21<p\le21<p≤2,

∣β^L−β∗∣pp≤16{1+3sm}2(p−1)s(Aσκ2(s,m,3)log⁡Mn)p.|\hat\beta_L-\beta^*|_p^p\le16\Big\{1+3\sqrt{\tfrac sm}\Big\}^{2(p-1)}s\Big(\frac{A\sigma}{\kappa^2(s,m,3)}\sqrt{\frac{\log M}{n}}\Big)^p .∣β^​L​−β∗∣pp​≤16{1+3ms​​}2(p−1)s(κ2(s,m,3)Aσ​nlogM​​)p.

Milestones

In the order in which the paper's proof uses them:

  1. (B.4): the noise event A=⋂j{2∣1nx(j)⊤w∣≤r}\mathcal A=\bigcap_j\{2|\tfrac1n x_{(j)}^\top w|\le r\}A=⋂j​{2∣n1​x(j)⊤​w∣≤r} has P(Ac)≤M1−A2/8\mathbb P(\mathcal A^c)\le M^{1-A^2/8}P(Ac)≤M1−A2/8.
  2. (B.6): the optimality conditions of the Lasso.
  3. Lemma B.1 (Section 7 case): the basic inequality (B.1) for all β\betaβ, the residual bound (B.2) and the sparsity bound M(β^L)≤4ϕmax⁡∥fβ^L−f∥n2/r2\mathcal M(\hat\beta_L)\le4\phi_{\max}\|f_{\hat\beta_L}-f\|_n^2/r^2M(β^​L​)≤4ϕmax​∥fβ^​L​​−f∥n2​/r2 (B.3).
  4. Corollary B.2: the error δ=β^L−β\delta=\hat\beta_L-\betaδ=β^​L​−β lies in the cone ∣δJ0c∣1≤3∣δJ0∣1|\delta_{J_0^c}|_1\le3|\delta_{J_0}|_1∣δJ0c​​∣1​≤3∣δJ0​​∣1​.
  5. (B.30)–(B.31): on A\mathcal AA, 1n∣Xδ∣22≤16r2s/κ2\frac1n|X\delta|_2^2\le16r^2s/\kappa^2n1​∣Xδ∣22​≤16r2s/κ2 and ∣δJ0∣2≤4rs/κ2|\delta_{J_0}|_2\le4r\sqrt s/\kappa^2∣δJ0​​∣2​≤4rs​/κ2.
  6. (B.27) and (B.28) with c0=3c_0=3c0​=3: ℓ1\ell_1ℓ1​ and ℓ2\ell_2ℓ2​ norms of a cone vector.
  7. The ℓp\ell_pℓp​ interpolation ∑ajp≤b12−pb2p−1\sum a_j^p\le b_1^{2-p}b_2^{p-1}∑ajp​≤b12−p​b2p−1​.

Significance

The result. Theorem 7.2 gives, for fixed nnn and MMM rather than asymptotically, the rate slog⁡M/ns\log M/nslogM/n for the prediction loss and slog⁡M/ns\sqrt{\log M/n}slogM/n​ for the ℓ1\ell_1ℓ1​ loss of the Lasso, under a condition on the design only (RE), with no assumption on how MMM compares with nnn. The dependence on MMM is only logarithmic, which is what makes the Lasso usable when M≫nM\gg nM≫n. Bound (7.9) shows that the Lasso selects at most a constant multiple of sss variables, and (7.10) covers every ℓp\ell_pℓp​ loss between ℓ1\ell_1ℓ1​ and ℓ2\ell_2ℓ2​. Together with Theorem 7.1 for the Dantzig selector, the result shows that the two estimators have the same rates.

Formalizing it. The theorem is proved on paper. As far as a search of the platform shows, there is no machine-checked proof of a probabilistic Lasso rate. The closest platform statement, HighDimStat.SparseLinear.lasso_l2_error_bound (Wainwright, Theorem 7.13(a)), is deterministic, assumes a lower bound on the regularisation parameter in place of Gaussian noise, uses a restricted eigenvalue condition over the cone of one fixed support, and concludes an ℓ2\ell_2ℓ2​ bound with a different constant. A formal proof of Theorem 7.2 would supply the Gaussian maximal inequality, the Lasso optimality conditions, and the cone and interpolation inequalities as reusable lemmas.

Difficulty

Each step is short on paper, and none of the steps is deep. The main work is in three places. First, the probability: the event on which the deterministic argument runs involves all MMM correlations 1nx(j)⊤w\frac1n x_{(j)}^\top wn1​x(j)⊤​w at once, and its probability must be bounded by exactly M1−A2/8M^{1-A^2/8}M1−A2/8, which requires the law of a linear combination of independent Gaussians and a sharp Gaussian tail estimate, not a generic concentration bound with unspecified constants. Second, the Lasso is defined only through its minimising property, while the sparsity bound (7.9) is a statement about the number of non-zero coordinates of a minimiser of a non-differentiable objective; the characterisation (B.6) of minimisers is not in Mathlib. Third, (7.10) involves two restricted eigenvalue constants, a ranking of coordinates with possible ties, and real exponents, and every constant has to come out exactly.

The obvious idea of proving (7.7)–(7.10) for one fixed minimiser does not suffice: the statement quantifies over every minimiser on a single event.

Formalization scope

The design XXX is a Matrix (Fin n) (Fin M) ℝ; vectors are functions Fin M → ℝ. The noise is a family W : Fin n → Ω → ℝ of independent, measurable random variables with law gaussianReal 0 σ² on a probability space, and y(ω)=Xβ∗+W(ω)y(\omega)=X\beta^*+W(\omega)y(ω)=Xβ∗+W(ω). The probabilistic conclusion is one measurable event EEE with P(E)≥1−M1−A2/8\mathbb P(E)\ge1-M^{1-A^2/8}P(E)≥1−M1−A2/8 on which every minimiser of (7.2) satisfies all bounds; the event does not depend on the minimiser, on mmm or on ppp. log⁡\loglog is the natural logarithm.

RE(s,3)(s,3)(s,3) and RE(s,m,3)(s,m,3)(s,m,3) are stated through witnesses: a predicate "κn∣δJ0∣2≤∣Xδ∣2\kappa\sqrt n|\delta_{J_0}|_2\le|X\delta|_2κn​∣δJ0​​∣2​≤∣Xδ∣2​ for every admissible J0J_0J0​ and δ\deltaδ", and the theorem holds for every witness κ>0\kappa>0κ>0. Because the paper's κ(s,c0)\kappa(s,c_0)κ(s,c0​) is an attained minimum, every witness is at most it and the bounds decrease in κ\kappaκ, so this is equivalent to the printed statement. The assumption quantifies over every J0J_0J0​ with ∣J0∣≤s|J_0|\le s∣J0​∣≤s, as on page 7, not only over the support of β∗\beta^*β∗. ϕmax⁡\phi_{\max}ϕmax​ is the supremum of 1n∣Xx∣22\frac1n|Xx|_2^2n1​∣Xx∣22​ over unit vectors xxx. Lemma B.1 is stated in its Section-7 specialisation (unit column norms, f=Xβ∗f=X\beta^*f=Xβ∗), the form used in the proof of Theorem 7.2; (B.28) is stated for every c0>0c_0>0c0​>0 and (B.27) likewise, since the paper writes them with c0=1c_0=1c0​=1 and invokes them with c0=3c_0=3c0​=3. The printed Theorem 7.2 needs no correction; all four constants were checked against the proof.

A formalization in which the noise is not Gaussian, the Lasso predicate can be vacuous, the RE condition is imposed only on the support of β∗\beta^*β∗, or the probability is that of a non-measurable set, is not this theorem and is ruled out by the statement.

Needed infrastructure: Gaussian tail bounds and the law of a linear combination of independent Gaussians (largely in Mathlib), subdifferential calculus for ℓ1\ell_1ℓ1​-penalised least squares, and elementary finite-sum inequalities. The cone inequalities (B.27)–(B.28), the interpolation inequality and the optimality conditions (B.6) are reusable in other sparse-estimation missions; contributions to any milestone are welcome.

Selected references

  • P. J. Bickel, Y. Ritov, A. B. Tsybakov, Simultaneous analysis of Lasso and Dantzig selector, Ann. Statist. 37(4), 1705–1732, 2009. arXiv:0801.1095v3, https://arxiv.org/abs/0801.1095 ; https://doi.org/10.1214/08-AOS620
  • F. Bunea, A. B. Tsybakov, M. H. Wegkamp, Sparsity oracle inequalities for the Lasso, Electron. J. Statist. 1, 169–194, 2007. https://doi.org/10.1214/07-EJS008
  • E. Candès, T. Tao, The Dantzig selector: statistical estimation when p is much larger than n, Ann. Statist. 35(6), 2313–2351, 2007. https://arxiv.org/abs/math/0506081
  • R. Tibshirani, Regression shrinkage and selection via the lasso, J. R. Statist. Soc. B 58(1), 267–288, 1996. https://doi.org/10.1111/j.2517-6161.1996.tb02080.x
  • M. J. Wainwright, High-Dimensional Statistics: A Non-Asymptotic Viewpoint, Cambridge University Press, 2019, Chapter 7. https://doi.org/10.1017/9781108627771
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OptimizationProbabilityStatistics·Captain: mikedeng1

Simultaneous Analysis of Lasso and Dantzig Selector IV: Estimation and Prediction Error Bounds for the Dantzig SelectorResearch Paper

Motivation

In high-dimensional linear regression the number of unknown coefficients MMM may be much larger than the number of observations nnn, and the coefficient vector can only be recovered because it is assumed to be sparse: few of its entries are non-zero. Two convex estimators dominate this setting: the Lasso of Tibshirani (1996), an ℓ1\ell_1ℓ1​-penalized least-squares estimator, and the Dantzig selector of Candès and Tao (2007), which minimizes the ℓ1\ell_1ℓ1​ norm subject to a bound on the correlation between the residual and the columns of the design. Both are used routinely in statistics, signal processing and machine learning, and their rates of convergence determine how many observations suffice to estimate a sparse vector.

Bickel, Ritov and Tsybakov (arXiv:0801.1095; Ann. Statist. 37(4), 2009) analysed the two estimators side by side under a single, weak condition on the design, the restricted eigenvalue (RE) assumption. This mission formalizes their rates for the Dantzig selector, Theorem 7.1 of the paper.

Timeline. Candès and Tao (Ann. Statist. 35, 2007) introduced the Dantzig selector and bounded its ℓ2\ell_2ℓ2​ error under a uniform uncertainty principle on the design. Bickel, Ritov and Tsybakov (2009) replaced that condition by the RE assumptions, which are implied by it (their Lemma 4.1), and obtained ℓp\ell_pℓp​ bounds for every 1≤p≤21\le p\le21≤p≤2 and a prediction bound, with explicit constants. Later work (van de Geer and Bühlmann, EJS 2009) compared RE with the compatibility condition and other design conditions.

Setting

Observations follow the linear model

y=Xβ∗+w,y=X\beta^*+w,y=Xβ∗+w,

where X∈Rn×MX\in\mathbb R^{n\times M}X∈Rn×M is a deterministic design matrix, n≥1n\ge1n≥1, M≥2M\ge2M≥2, β∗∈RM\beta^*\in\mathbb R^Mβ∗∈RM is unknown, and w=(W1,…,Wn)w=(W_1,\dots,W_n)w=(W1​,…,Wn​) has independent N(0,σ2)\mathcal N(0,\sigma^2)N(0,σ2) coordinates with σ>0\sigma>0σ>0. The columns are normalized: every diagonal element of the Gram matrix XTX/nX^TX/nXTX/n equals 1.

For β∈RM\beta\in\mathbb R^Mβ∈RM, J(β)={j:βj≠0}J(\beta)=\{j:\beta_j\ne0\}J(β)={j:βj​=0} is its support and M(β)=∣J(β)∣\mathcal M(\beta)=|J(\beta)|M(β)=∣J(β)∣ its sparsity; β∗\beta^*β∗ satisfies M(β∗)≤s\mathcal M(\beta^*)\le sM(β∗)≤s for an integer 1≤s≤M1\le s\le M1≤s≤M. Norms are ∣δ∣p=(∑j∣δj∣p)1/p|\delta|_p=(\sum_j|\delta_j|^p)^{1/p}∣δ∣p​=(∑j​∣δj​∣p)1/p and ∣v∣22=∑ivi2|v|_2^2=\sum_iv_i^2∣v∣22​=∑i​vi2​; for an index set JJJ, δJ\delta_JδJ​ keeps the coordinates of δ\deltaδ in JJJ and sets the others to 0, and JcJ^cJc is the complement of JJJ.

With a tuning level r=Aσlog⁡M/nr=A\sigma\sqrt{\log M/n}r=AσlogM/n​, A>2A>\sqrt2A>2​, the Dantzig selector is any minimizer

β^D∈arg⁡min⁡β∈Λ∣β∣1,Λ={β∈RM: ∣1nXT(y−Xβ)∣∞≤r}.\hat\beta_D\in\arg\min_{\beta\in\Lambda}|\beta|_1,\qquad \Lambda=\Big\{\beta\in\mathbb R^M:\ \Big|\tfrac1nX^T(y-X\beta)\Big|_\infty\le r\Big\}.β^​D​∈argβ∈Λmin​∣β∣1​,Λ={β∈RM: ​n1​XT(y−Xβ)​∞​≤r}.

The cone condition at an index set J0J_0J0​ with constant c0>0c_0>0c0​>0 is ∣δJ0c∣1≤c0∣δJ0∣1|\delta_{J_0^c}|_1\le c_0|\delta_{J_0}|_1∣δJ0c​​∣1​≤c0​∣δJ0​​∣1​. Assumption RE(s,c0)(s,c_0)(s,c0​) asks that

κ(s,c0)=min⁡∣J0∣≤s min⁡δ≠0, ∣δJ0c∣1≤c0∣δJ0∣1∣Xδ∣2n ∣δJ0∣2>0.\kappa(s,c_0)=\min_{|J_0|\le s}\ \min_{\delta\ne0,\ |\delta_{J_0^c}|_1\le c_0|\delta_{J_0}|_1}\frac{|X\delta|_2}{\sqrt n\,|\delta_{J_0}|_2}>0 .κ(s,c0​)=∣J0​∣≤smin​ δ=0, ∣δJ0c​​∣1​≤c0​∣δJ0​​∣1​min​n​∣δJ0​​∣2​∣Xδ∣2​​>0.

Assumption RE(s,m,c0)(s,m,c_0)(s,m,c0​) is the same with ∣δJ01∣2|\delta_{J_{01}}|_2∣δJ01​​∣2​ in the denominator, where J01=J0∪J1J_{01}=J_0\cup J_1J01​=J0​∪J1​ and J1J_1J1​ collects the mmm largest ∣δj∣|\delta_j|∣δj​∣ outside J0J_0J0​; it is used for s≤ms\le ms≤m, s+m≤Ms+m\le Ms+m≤M.

Formalization targets

Goal: Theorem 7.1

With probability at least 1−M1−A2/21-M^{1-A^2/2}1−M1−A2/2, every Dantzig selector satisfies

∣β^D−β∗∣1≤8Aκ2(s,1) σslog⁡Mn,∣X(β^D−β∗)∣22≤16A2κ2(s,1) σ2slog⁡M,|\hat\beta_D-\beta^*|_1\le\frac{8A}{\kappa^2(s,1)}\,\sigma s\sqrt{\frac{\log M}{n}},\qquad |X(\hat\beta_D-\beta^*)|_2^2\le\frac{16A^2}{\kappa^2(s,1)}\,\sigma^2s\log M,∣β^​D​−β∗∣1​≤κ2(s,1)8A​σsnlogM​​,∣X(β^​D​−β∗)∣22​≤κ2(s,1)16A2​σ2slogM,

and, on the same event, if RE(s,m,1)(s,m,1)(s,m,1) holds, simultaneously for all 1<p≤21<p\le21<p≤2,

∣β^D−β∗∣pp≤2p−1 8{1+sm}2(p−1)s(Aσκ2(s,m,1)log⁡Mn)p.|\hat\beta_D-\beta^*|_p^p\le2^{p-1}\,8\Big\{1+\sqrt{\tfrac sm}\Big\}^{2(p-1)}s\Big(\frac{A\sigma}{\kappa^2(s,m,1)}\sqrt{\frac{\log M}{n}}\Big)^p .∣β^​D​−β∗∣pp​≤2p−18{1+ms​​}2(p−1)s(κ2(s,m,1)Aσ​nlogM​​)p.

Milestones

In the order the proof of the paper uses them:

  1. The noise event B=⋂j{∣1n∑iXijWi∣≤r∥fj∥n}\mathcal B=\bigcap_j\{|\frac1n\sum_iX_{ij}W_i|\le r\|f_j\|_n\}B=⋂j​{∣n1​∑i​Xij​Wi​∣≤r∥fj​∥n​} has P{Bc}≤M1−A2/2\mathbb P\{\mathcal B^c\}\le M^{1-A^2/2}P{Bc}≤M1−A2/2 (proof of Lemma B.3).
  2. Lemma B.3, (B.9): for any β\betaβ satisfying the Dantzig constraint, δ=β^D−β\delta=\hat\beta_D-\betaδ=β^​D​−β satisfies the cone condition at J(β)J(\beta)J(β) with c0=1c_0=1c0​=1.
  3. (B.25): on B\mathcal BB, β∗∈Λ\beta^*\in\Lambdaβ∗∈Λ, 1n∣XTXδ∣∞≤2r\frac1n|X^TX\delta|_\infty\le2rn1​∣XTXδ∣∞​≤2r, and 1n∣Xδ∣22≤4rs ∣δJ0∣2\frac1n|X\delta|_2^2\le4r\sqrt s\,|\delta_{J_0}|_2n1​∣Xδ∣22​≤4rs​∣δJ0​​∣2​.
  4. (B.26): under RE(s,1)(s,1)(s,1), 1n∣Xδ∣22≤16r2s/κ2\frac1n|X\delta|_2^2\le16r^2s/\kappa^2n1​∣Xδ∣22​≤16r2s/κ2 and ∣δJ0∣2≤4rs/κ2|\delta_{J_0}|_2\le4r\sqrt s/\kappa^2∣δJ0​​∣2​≤4rs​/κ2.
  5. (B.27): on the cone, ∣δ∣1≤(1+c0)s ∣δJ0∣2|\delta|_1\le(1+c_0)\sqrt s\,|\delta_{J_0}|_2∣δ∣1​≤(1+c0​)s​∣δJ0​​∣2​.
  6. (B.28): on the cone, ∣δ∣2≤(1+c0s/m) ∣δJ01∣2|\delta|_2\le(1+c_0\sqrt{s/m})\,|\delta_{J_{01}}|_2∣δ∣2​≤(1+c0​s/m​)∣δJ01​​∣2​.
  7. (B.29): under RE(s,m,1)(s,m,1)(s,m,1), ∣δ∣22≤16(1+s/m)2(rs/κ2)2|\delta|_2^2\le16(1+\sqrt{s/m})^2(r\sqrt s/\kappa^2)^2∣δ∣22​≤16(1+s/m​)2(rs​/κ2)2.
  8. Interpolation: ∑jaj≤b1\sum_ja_j\le b_1∑j​aj​≤b1​, ∑jaj2≤b2\sum_ja_j^2\le b_2∑j​aj2​≤b2​, aj≥0a_j\ge0aj​≥0 imply ∑jajp≤b12−pb2p−1\sum_ja_j^p\le b_1^{2-p}b_2^{p-1}∑j​ajp​≤b12−p​b2p−1​ for 1<p≤21<p\le21<p≤2.

Significance

The result. Theorem 7.1 shows that, up to the factor log⁡M\log MlogM, the Dantzig selector estimates an sss-sparse vector as well as least squares would if the support were known: the prediction error 1n∣X(β^D−β∗)∣22\frac1n|X(\hat\beta_D-\beta^*)|_2^2n1​∣X(β^​D​−β∗)∣22​ is of order σ2slog⁡M/n\sigma^2s\log M/nσ2slogM/n, and the ℓp\ell_pℓp​ errors are of order s1/pσlog⁡M/ns^{1/p}\sigma\sqrt{\log M/n}s1/pσlogM/n​. The bounds hold for any MMM, including M≫nM\gg nM≫n, provided only that RE holds, and every constant is explicit. The paper's Theorem 7.2 gives the same rates for the Lasso; comparing the two is the paper's main message.

Formalizing it. The theorem is proved in the paper; to the best of current knowledge it has not been machine-checked. A complete formal proof would provide: a verified Gaussian maximal inequality for the noise event, the deterministic cone and RE arithmetic that underlies essentially all ℓ1\ell_1ℓ1​-regularized estimation theory, and a reusable ℓ1\ell_1ℓ1​–ℓ2\ell_2ℓ2​ interpolation lemma. Most milestones are deterministic and independent of the probability layer.

Difficulty

The obvious argument — compare β^D\hat\beta_Dβ^​D​ with β∗\beta^*β∗ in Euclidean norm using the smallest eigenvalue of XTX/nX^TX/nXTX/n — fails because that eigenvalue is 0 whenever M>nM>nM>n. The proof must instead show that the error vector lies in a cone on which XXX is injective in a quantitative sense, and this uses the optimality of β^D\hat\beta_Dβ^​D​ (not just feasibility) together with the event B\mathcal BB on which β∗\beta^*β∗ itself is feasible. The ℓp\ell_pℓp​ bound needs a second, stronger condition RE(s,m,1)(s,m,1)(s,m,1) and a control of the tail of the error outside the mmm largest coordinates. On the formal side, handling the non-uniqueness of the minimizer, real powers with exponent p−1p-1p−1 or 2−p2-p2−p, and the union over MMM Gaussian tails with the exact constant M1−A2/2M^{1-A^2/2}M1−A2/2 all need care.

Formalization scope

Vectors are functions Fin M → ℝ, the design is Matrix (Fin n) (Fin M) ℝ; the paper's dictionary of functions enters only through XXX. The unit diagonal of XTX/nX^TX/nXTX/n is a hypothesis, not a normalization performed in the proof. The noise is W : Fin n → Ω → ℝ on a probability space, measurable, mutually independent, each of law N(0,σ2)\mathcal N(0,\sigma^2)N(0,σ2); log⁡\loglog is the natural logarithm. The Dantzig selector is a predicate (feasible and of minimal ℓ1\ell_1ℓ1​ norm among feasible vectors), and every result is stated for every minimizer. RE(s,c0)(s,c_0)(s,c0​) and RE(s,m,c0)(s,m,c_0)(s,m,c0​) are stated through a witness κ\kappaκ (a number with the defining lower-bound property); κ(s,c0)\kappa(s,c_0)κ(s,c0​) is the largest witness and the bounds decrease in κ\kappaκ, so the statements are equivalent to the paper's while avoiding the value of a real infimum over an empty set. Two witnesses are kept apart: κ\kappaκ for RE(s,1)(s,1)(s,1) in (7.4)–(7.5), κ′\kappa'κ′ for RE(s,m,1)(s,m,1)(s,m,1) in (7.6). Ties in the choice of the mmm largest coordinates are handled by quantifying over every admissible J1J_1J1​. The probability statement asserts one measurable event EEE with P(E)≥1−M1−A2/2\mathbb P(E)\ge1-M^{1-A^2/2}P(E)≥1−M1−A2/2 on which all three bounds hold for every minimizer, every admissible mmm, every witness κ′\kappa'κ′ and every ppp.

The event EEE is fixed before the minimizer is quantified, so a formalization in which the event depends on β^D\hat\beta_Dβ^​D​, or in which RE is a hypothesis about the random error vector rather than the design, would be a different (weaker) statement and is not accepted. The deterministic milestones (B.26)–(B.29) take the conclusion of (B.25) as a hypothesis; they are true for every vector satisfying their hypotheses and are not restricted to the event.

A complete development needs Gaussian tail bounds and a union bound (Mathlib's gaussianReal), finite Hölder-type inequalities for real exponents, and elementary sorting arguments for the tail outside J01J_{01}J01​. The cone, RE and interpolation lemmas are reusable for the Lasso (Theorem 7.2, a sister mission) and beyond. Proofs of any milestone are welcome independently.

Selected references

  • P. J. Bickel, Y. Ritov, A. B. Tsybakov, Simultaneous analysis of Lasso and Dantzig selector, Ann. Statist. 37(4), 1705–1732, 2009. arXiv:0801.1095v3: https://arxiv.org/abs/0801.1095
  • E. Candès, T. Tao, The Dantzig selector: statistical estimation when p is much larger than n, Ann. Statist. 35(6), 2313–2351, 2007. https://doi.org/10.1214/009053606000001523
  • R. Tibshirani, Regression shrinkage and selection via the lasso, J. R. Stat. Soc. B 58(1), 267–288, 1996. https://doi.org/10.1111/j.2517-6161.1996.tb02080.x
  • S. van de Geer, P. Bühlmann, On the conditions used to prove oracle results for the Lasso, Electron. J. Statist. 3, 1360–1392, 2009. https://doi.org/10.1214/09-EJS506
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OptimizationProbabilityStatistics·Captain: mikedeng1

Simultaneous Analysis of Lasso and Dantzig Selector II: Approximate Equivalence of the Lasso and Dantzig Prediction LossesResearch Paper

Motivation

Two estimators dominate sparse high-dimensional regression, where the number MMM of candidate regressors can far exceed the sample size nnn. The Lasso (Tibshirani, 1996) minimises a least-squares criterion plus an ℓ1\ell_1ℓ1​ penalty. The Dantzig selector (Candès and Tao, 2007) minimises the ℓ1\ell_1ℓ1​ norm of the coefficients subject to a bound on the correlation between the residual and the regressors, and is computed by a linear program. They were proposed independently and first analysed under different assumptions: sparsity oracle inequalities for the Lasso (Bunea, Tsybakov and Wegkamp, 2007) and ℓ2\ell_2ℓ2​ bounds for the Dantzig selector under a uniform uncertainty principle (Candès and Tao, 2007). A practitioner choosing between them needs to know whether guarantees for one say anything about the other.

Bickel, Ritov and Tsybakov (arXiv:0801.1095; Ann. Statist. 37(4), 2009, doi:10.1214/08-AOS620) analyse both estimators in parallel under one assumption on the design, the restricted eigenvalue condition. Their main message is that, under sparsity, the two estimators "exhibit similar behavior" (p. 2). Section 5 makes this precise: the prediction losses of the two estimators are close. The result holds in a nonparametric model: the regression function need not be a combination of the regressors. This mission formalizes that comparison, Theorem 5.1 of the paper.

Setting

Let f1,…,fMf_1,\dots,f_Mf1​,…,fM​ be real functions (the dictionary) on a set Z\mathcal ZZ, and Z1,…,Zn∈ZZ_1,\dots,Z_n\in\mathcal ZZ1​,…,Zn​∈Z fixed design points, with n≥1n\ge1n≥1 and M≥2M\ge2M≥2. The design matrix is X=(fj(Zi))∈Rn×MX=(f_j(Z_i))\in\mathbb R^{n\times M}X=(fj​(Zi​))∈Rn×M. Observations are Yi=f(Zi)+WiY_i=f(Z_i)+W_iYi​=f(Zi​)+Wi​, where fff is an unknown function and W1,…,WnW_1,\dots,W_nW1​,…,Wn​ are independent N(0,σ2)\mathcal N(0,\sigma^2)N(0,σ2) with σ>0\sigma>0σ>0. Write y=(Yi)y=(Y_i)y=(Yi​), f=(f(Zi))\boldsymbol f=(f(Z_i))f=(f(Zi​)) and w=(Wi)w=(W_i)w=(Wi​), so y=f+wy=\boldsymbol f+wy=f+w.

The empirical norm of ggg is ∥g∥n=(1n∑ig(Zi)2)1/2\|g\|_n=(\tfrac1n\sum_i g(Z_i)^2)^{1/2}∥g∥n​=(n1​∑i​g(Zi​)2)1/2. Every column has ∥fj∥n≠0\|f_j\|_n\neq0∥fj​∥n​=0, and fmax⁡=max⁡j∥fj∥nf_{\max}=\max_j\|f_j\|_nfmax​=maxj​∥fj​∥n​. For β∈RM\beta\in\mathbb R^Mβ∈RM, fβ=∑jβjfjf_\beta=\sum_j\beta_jf_jfβ​=∑j​βj​fj​ has value vector XβX\betaXβ. The support is J(β)={j:βj≠0}J(\beta)=\{j:\beta_j\ne0\}J(β)={j:βj​=0} and the sparsity is M(β)=∣J(β)∣\mathcal M(\beta)=|J(\beta)|M(β)=∣J(β)∣. For J⊆{1,…,M}J\subseteq\{1,\dots,M\}J⊆{1,…,M}, δJ\delta_JδJ​ agrees with δ\deltaδ on JJJ and vanishes elsewhere.

Fix r>0r>0r>0. The Lasso β^L\hat\beta_Lβ^​L​ is any minimiser of

1n∑i=1n(Yi−fβ(Zi))2+2r∑j=1M∥fj∥n∣βj∣.\frac1n\sum_{i=1}^n\big(Y_i-f_\beta(Z_i)\big)^2+2r\sum_{j=1}^M\|f_j\|_n|\beta_j| .n1​i=1∑n​(Yi​−fβ​(Zi​))2+2rj=1∑M​∥fj​∥n​∣βj​∣.

With D=diag(∥f1∥n2,…,∥fM∥n2)D=\mathrm{diag}(\|f_1\|_n^2,\dots,\|f_M\|_n^2)D=diag(∥f1​∥n2​,…,∥fM​∥n2​), the Dantzig constraint is ∣1nD−1/2X⊤(y−Xβ)∣∞≤r|\tfrac1nD^{-1/2}X^\top(y-X\beta)|_\infty\le r∣n1​D−1/2X⊤(y−Xβ)∣∞​≤r. The Dantzig selector β^D\hat\beta_Dβ^​D​ is any vector of smallest ∣β∣1=∑j∣βj∣|\beta|_1=\sum_j|\beta_j|∣β∣1​=∑j​∣βj​∣ that satisfies it. The estimators are f^L=fβ^L\hat f_L=f_{\hat\beta_L}f^​L​=fβ^​L​​ and f^D=fβ^D\hat f_D=f_{\hat\beta_D}f^​D​=fβ^​D​​.

Assumption RE(s,c0)(s,c_0)(s,c0​) with 1≤s≤M1\le s\le M1≤s≤M, c0>0c_0>0c0​>0 asks that

κ(s,c0)=min⁡∣J0∣≤s min⁡δ≠0, ∣δJ0c∣1≤c0∣δJ0∣1 ∣Xδ∣2n ∣δJ0∣2>0.\kappa(s,c_0)=\min_{|J_0|\le s}\ \min_{\delta\ne0,\ |\delta_{J_0^c}|_1\le c_0|\delta_{J_0}|_1}\ \frac{|X\delta|_2}{\sqrt n\,|\delta_{J_0}|_2}>0 .κ(s,c0​)=∣J0​∣≤smin​ δ=0, ∣δJ0c​​∣1​≤c0​∣δJ0​​∣1​min​ n​∣δJ0​​∣2​∣Xδ∣2​​>0.

Throughout, r=Aσlog⁡M/nr=A\sigma\sqrt{\log M/n}r=AσlogM/n​, where log⁡\loglog is the natural logarithm.

Formalization targets

Goal: Theorem 5.1

Assume RE(s,1)(s,1)(s,1) with 1≤s≤M1\le s\le M1≤s≤M, and let A>22A>2\sqrt2A>22​. With probability at least 1−M1−A2/81-M^{1-A^2/8}1−M1−A2/8, every Lasso solution with M(β^L)≤s\mathcal M(\hat\beta_L)\le sM(β^​L​)≤s and every Dantzig selector satisfy

∣ ∥f^D−f∥n2−∥f^L−f∥n2 ∣≤16A2 M(β^L)σ2n fmax⁡2κ2(s,1) log⁡M.\Big|\,\|\hat f_D-f\|_n^2-\|\hat f_L-f\|_n^2\,\Big|\le16A^2\,\frac{\mathcal M(\hat\beta_L)\sigma^2}{n}\,\frac{f_{\max}^2}{\kappa^2(s,1)}\,\log M .​∥f^​D​−f∥n2​−∥f^​L​−f∥n2​​≤16A2nM(β^​L​)σ2​κ2(s,1)fmax2​​logM.

Milestones

The proof uses one probabilistic event and two one-sided deterministic inequalities.

  1. The Lasso satisfies the Dantzig constraint (2.3).
  2. The noise event A=⋂j{2∣1n∑iXijWi∣≤r∥fj∥n}\mathcal A=\bigcap_j\{2|\tfrac1n\sum_iX_{ij}W_i|\le r\|f_j\|_n\}A=⋂j​{2∣n1​∑i​Xij​Wi​∣≤r∥fj​∥n​} has P(Ac)≤M1−A2/8\mathbb P(\mathcal A^c)\le M^{1-A^2/8}P(Ac)≤M1−A2/8 (B.4).
  3. On A\mathcal AA, ∣1nX⊤(f−Xβ^L)∣∞≤3rfmax⁡/2|\tfrac1nX^\top(\boldsymbol f-X\hat\beta_L)|_\infty\le 3rf_{\max}/2∣n1​X⊤(f−Xβ^​L​)∣∞​≤3rfmax​/2 (Lemma B.1, (B.2)).
  4. The Dantzig error lies in the cone ∣δJ0c∣1≤∣δJ0∣1|\delta_{J_0^c}|_1\le|\delta_{J_0}|_1∣δJ0c​​∣1​≤∣δJ0​​∣1​ (Lemma B.3, (B.9)).
  5. On the larger event B⊇A\mathcal B\supseteq\mathcal AB⊇A, ∣1nX⊤(f−Xβ^D)∣∞≤2rfmax⁡|\tfrac1nX^\top(\boldsymbol f-X\hat\beta_D)|_\infty\le 2rf_{\max}∣n1​X⊤(f−Xβ^​D​)∣∞​≤2rfmax​ (Lemma B.3, (B.10)).
  6. ∥f^D−f∥n2≤∥f^L−f∥n2+16fmax⁡2r2M(β^L)/κ2\|\hat f_D-f\|_n^2\le\|\hat f_L-f\|_n^2+16f_{\max}^2r^2\mathcal M(\hat\beta_L)/\kappa^2∥f^​D​−f∥n2​≤∥f^​L​−f∥n2​+16fmax2​r2M(β^​L​)/κ2 on B\mathcal BB (B.15).
  7. ∥f^L−f∥n2≤∥f^D−f∥n2+9fmax⁡2r2M(β^L)/κ2\|\hat f_L-f\|_n^2\le\|\hat f_D-f\|_n^2+9f_{\max}^2r^2\mathcal M(\hat\beta_L)/\kappa^2∥f^​L​−f∥n2​≤∥f^​D​−f∥n2​+9fmax2​r2M(β^​L​)/κ2 on A\mathcal AA (B.17).

Further result: Theorem 5.2

Assume ∥fj∥n=1\|f_j\|_n=1∥fj​∥n​=1 for all jjj and RE(s,5)(s,5)(s,5). With probability at least 1−M1−A2/81-M^{1-A^2/8}1−M1−A2/8, whenever M(β^D)≤s\mathcal M(\hat\beta_D)\le sM(β^​D​)≤s,

∥f^L−f∥n2≤10∥f^D−f∥n2+81A2 M(β^D)σ2n log⁡Mκ2(s,5).\|\hat f_L-f\|_n^2\le10\|\hat f_D-f\|_n^2+81A^2\,\frac{\mathcal M(\hat\beta_D)\sigma^2}{n}\,\frac{\log M}{\kappa^2(s,5)} .∥f^​L​−f∥n2​≤10∥f^​D​−f∥n2​+81A2nM(β^​D​)σ2​κ2(s,5)logM​.

Significance

The result. Theorem 5.1 bounds the gap between the two prediction losses by the rate M(β^L)σ2log⁡M/n\mathcal M(\hat\beta_L)\sigma^2\log M/nM(β^​L​)σ2logM/n of a sparse regression with M(β^L)\mathcal M(\hat\beta_L)M(β^​L​) parameters. The bound carries a factor fmax⁡2/κ2(s,1)f^2_{\max}/\kappa^2(s,1)fmax2​/κ2(s,1) that measures how ill-conditioned the Gram matrix is on sparse vectors. A prediction bound for one estimator therefore transfers to the other at this cost. The paper uses this transfer in Proposition 6.3, which combines Theorem 5.1 with the Lasso oracle inequality of Section 6 to derive an oracle inequality for the Dantzig selector. The theorem requires no assumption relating fff to the dictionary.

Formalizing it. The result has been proved since 2009, and this mission formalizes that proof. None of the objects involved exists on Prove2Me yet: the weighted Lasso, the Dantzig selector and the Gaussian noise events. The Wainwright series on the platform defines a differently normalised Lasso with an unweighted penalty, a single fixed support and a different restricted eigenvalue condition, so it cannot be reused here. A machine-checked proof would also confirm the paper's constants, 16A216A^216A2 and the thresholds 222\sqrt222​ and M1−A2/8M^{1-A^2/8}M1−A2/8, which appear in all later analyses.

Difficulty

Each estimator is defined only implicitly, as the solution of an optimisation problem, and neither need be unique. Comparing their losses directly gives ±2nδ⊤X⊤(f−Xβ^)\pm\tfrac2n\delta^\top X^\top(\boldsymbol f-X\hat\beta)±n2​δ⊤X⊤(f−Xβ^​) plus 1n∣Xδ∣22\tfrac1n|X\delta|_2^2n1​∣Xδ∣22​ with δ=β^L−β^D\delta=\hat\beta_L-\hat\beta_Dδ=β^​L​−β^​D​. A crude bound on the cross term, ∣δ∣1⋅∣X⊤(⋅)∣∞|\delta|_1\cdot|X^\top(\cdot)|_\infty∣δ∣1​⋅∣X⊤(⋅)∣∞​, yields an error proportional to ∣δ∣1|\delta|_1∣δ∣1​. This does not produce the sparse rate unless ∣δ∣1|\delta|_1∣δ∣1​ is controlled by ∣Xδ∣2|X\delta|_2∣Xδ∣2​. That control needs δ\deltaδ to lie in the restricted eigenvalue cone at the support of the random, data-dependent vector β^L\hat\beta_Lβ^​L​. The restricted eigenvalue condition must therefore hold uniformly over supports of size at most sss; a condition for one fixed support does not suffice. The probabilistic part is a union bound over MMM Gaussian coordinates, and it must be arranged so that a single event serves every minimiser of both programs.

Formalization scope

The dictionary and the design points enter every statement only through XXX and f\boldsymbol ff, so the Lean statements take X : Matrix (Fin n) (Fin M) ℝ and f : Fin n → ℝ directly, and fff is arbitrary. The noise is a family W : Fin n → Ω → ℝ of measurable, mutually independent random variables with law gaussianReal 0 σ², and y=f+W(ω)y=f+W(\omega)y=f+W(ω). The Lasso and the Dantzig selector are predicates (IsLasso, IsDantzig), and every theorem is stated for every solution. The Dantzig constraint is written coordinatewise as ∣1n∑iXij(yi−(Xβ)i)∣≤r∥fj∥n|\tfrac1n\sum_iX_{ij}(y_i-(X\beta)_i)|\le r\|f_j\|_n∣n1​∑i​Xij​(yi​−(Xβ)i​)∣≤r∥fj​∥n​. The Lasso penalty and the Dantzig constraint are weighted by ∥fj∥n\|f_j\|_n∥fj​∥n​, and the Dantzig objective ∣β∣1|\beta|_1∣β∣1​ is unweighted, exactly as in the paper. Theorem 5.1 does not normalise the columns.

RE(s,c0)(s,c_0)(s,c0​) is stated through a witness: a real κ>0\kappa>0κ>0 with κn∣δJ0∣2≤∣Xδ∣2\kappa\sqrt n|\delta_{J_0}|_2\le|X\delta|_2κn​∣δJ0​​∣2​≤∣Xδ∣2​ on the cone, for all ∣J0∣≤s|J_0|\le s∣J0​∣≤s. The paper's κ(s,c0)\kappa(s,c_0)κ(s,c0​) is attained, so it is the largest witness. Every bound decreases in κ\kappaκ, so this reading is equivalent to the paper's and avoids a real infimum over an empty set. "With probability at least ppp" becomes the existence of a measurable event EEE with P(E)≥p\mathbb P(E)\ge pP(E)≥p on which the conclusion holds for every Lasso solution and every Dantzig selector. The condition M(β^L)≤s\mathcal M(\hat\beta_L)\le sM(β^​L​)≤s is imposed inside the event, per realisation. The milestones (B.2), (B.10), (B.15) and (B.17) are stated deterministically, on the noise events A\mathcal AA and B\mathcal BB as predicates on the noise vector; this is how the proof uses them. (B.4) is stated for every A>0A>0A>0, which is stronger than the paper's A>22A>2\sqrt2A>22​ and still true.

The goal cannot be made vacuous. For A>22A>2\sqrt2A>22​ the probability bound 1−M1−A2/81-M^{1-A^2/8}1−M1−A2/8 is positive. Lasso solutions exist because r>0r>0r>0 and every ∥fj∥n>0\|f_j\|_n>0∥fj​∥n​>0, and Dantzig selectors exist because the Lasso is feasible. RE(s,1)(s,1)(s,1) with κ=1\kappa=1κ=1 holds for X=n IX=\sqrt n\,IX=n​I.

A complete development needs:

  • subgradient optimality for the weighted Lasso;
  • a Gaussian tail bound P(∣η∣≥t)≤e−t2/2\mathbb P(|\eta|\ge t)\le e^{-t^2/2}P(∣η∣≥t)≤e−t2/2 together with the law of a weighted sum of independent Gaussians;
  • Cauchy–Schwarz on supports;
  • the quadratic bound bx−x2≤b2/4bx-x^2\le b^2/4bx−x2≤b2/4.

The noise-event lemmas and the Lasso optimality condition can be reused by the companion missions on this paper. Proofs of any milestone are welcome, including proofs that route (B.4) through Mathlib's sub-Gaussian API.

Selected references

  • P. J. Bickel, Y. Ritov, A. B. Tsybakov, Simultaneous analysis of Lasso and Dantzig selector, Ann. Statist. 37(4), 1705–1732, 2009. arXiv:0801.1095v3: https://arxiv.org/abs/0801.1095 ; doi:10.1214/08-AOS620
  • E. Candès, T. Tao, The Dantzig selector: statistical estimation when p is much larger than n, Ann. Statist. 35(6), 2313–2351, 2007. https://doi.org/10.1214/009053606000001523
  • R. Tibshirani, Regression shrinkage and selection via the lasso, J. R. Stat. Soc. B 58(1), 267–288, 1996. https://doi.org/10.1111/j.2517-6161.1996.tb02080.x
  • F. Bunea, A. B. Tsybakov, M. H. Wegkamp, Sparsity oracle inequalities for the Lasso, Electron. J. Stat. 1, 169–194, 2007. https://doi.org/10.1214/07-EJS008
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