Regularized Estimation of High Dimensional Covariance Matrices. Peter Bickel. January, 2008

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1 Regularized Estimation of High Dimensional Covariance Matrices Peter Bickel Cambridge January, 2008 With Thanks to E. Levina (Joint collaboration, slides) I. M. Johnstone (Slides) Choongsoon Bae (Slides)

2 Outline 1. Why estimate covariance matrices? 2. Some examples 3. Pathologies of empirical covariance matrix 4. Sparsity in various guises 5. A brief review of methods 6. Asymptotic theory for banding, thresholding, SPICE (LASSO penalized Gaussian log likelihood) 7. Simulations and data 8. Some future directions

3 Model X 1,..., X n : p vectors, I.I.D. F. Base model: F = N p (µ, Σ) More generally: E F X 2 < µ(f ) = E F X Σ(F ) = E F (X µ)(x µ) T Asymptotic theory: n, p, Σ T p,n Extensions to Stationary X 1,..., X n, Wei Biao Wu (2007).

4 Why Estimate Covariance Matrix? Principal component analysis (PCA) Linear or quadratic discriminant analysis (LDA/QDA) Inferring independence and conditional independence in Gaussian models(graphical models) Inference about the mean (e.g. longitudinal mean response curve) Covariance itself is usually not the end goal: PCA: Estimation of the eigenstructure LDA/QDA, Graphical models: Estimation of inverses

5 Covariance Matrices in Climate Research Example 1 Data courtesy of Serge Guillas Empirical Orthogonal Functions(EOF) aka principal components. X 1,..., X n : p dimensional stationary vector time series. Monthly Jan Temp Latitude longitude calls 5 o 5 o grid p=2592 n=157

6 Example 1 EOF pattern #1 (clim.pact)

7 Example 1 X k = {X (i, j) : i = latitude, j = longitude} X n p = ( X T 1,..., XT n E(X) = 0 Σ = Var(X) = E ) T ( XX T ) = p λ j e j e T j j=1 e 1,..., e p : Principal components. λ 1 > > λ p : Eigenvalues. Goal : Estimate, interpret e j, j = 1,..., K such that K j=1 λ j p j=1 λ j large.

8 Covariance Matrices in Climate Research Example 2 Data assimilation X j = ave. pressure, temperature,... in 50km 50km variable block of atmosphere, J 10 7, computer model. X i = X(t i ) i = 1,..., T. Y(t i ) : Data vectors. Ensemble: X U j (t), 1 j n. Data assimilated : X F j (t), 1 j n uses ˆΣ 1 as estimate of true [ Var(X U 1 )] 1 for Kalman gain.

9 Pathologies of Empirical Covariance Matrix Observe X 1,..., X n, i.i.d. p-variate random variables ˆΣ = 1 n n ( Xi X ) ( X i X ) T i=1 MLE, for Gaussian unbiased (almost), well-behaved (and well studied) for fixed p, n. But very noisy if p is large. Singular if p > n, so ˆΣ 1 is not uniquely defined. Computational issues with ˆΣ 1 for large p. LDA completely breaks down if p/n Eigenstructure can be inconsistent if p/n γ > 0.

10 Eigenvalues Description of spreading phenomenon: Empirical distribution function: for eigenvalues {ˆl i } p j=1 G p (t) = p 1 #{ˆl j t} G(t) g(t)dt. Marčenko-Pastur, (67), For F = N p (0, I ), p/n γ 1.5 n=4p n=p g MP (b+ t)(t b ) (t) =, 2πγt b ± = (1 ± γ)

11 Eigenvectors D.Paul (2006) For N p (0, Σ) λ Σ = diag(λ, 1,..., 1) = Eigenvector ê ˆλ, e λ,, 1 < λ < 1 + γ, e ê 2 a.s. 2

12 Fisher s LDA For classification into two populations using: X 1,..., X n N p (, Σ), X n+1,..., X 2n N p (, Σ) Minimax behaviour over T p,n = { Σ : T Σ 1 c > 0 } of LDA based on MLE Σ 1 (Moore-Penrose for p 2 n 2 ) is equivalent to coin tossing if p n (Bickel - Levina (2004)) True for any rule!. What if T consists of sparse matrices?

13 Eigenstructure Sparsity Write, Σ = Σ 0 + σ 2 I p, where Σ 0 = M λ j θ j θ t j j=1 and λ 1... λ M > 0, {θ j } orthonormal. Equivalent : M X i = µ+ λj v ji θ j +σz i, i = 1,..., n, v ji i.i.d. N (0, 1) j=1

14 Eigenstructure Sparsity defined p, n large but, (i) M small fixed. (ii) θ j sparse well approximated by θ js, θ js 0 s, s small. v js 0 # of nonzero coordinates of v.

15 Label Dependent Sparsity of Σ Write Σ = σ ij, σ ij small (effectively 0) for i j large More generally, given metric m on J, σ ij small if m(i, j) large Note: σ ij = 0 X i X j under Gaussianity.

16 Label Dependent Sparsity of Σ 1 If Σ 1 = σ ij, σ ij small for many (i, j) pairs if i j is large. Note: σ ij = 0 X i X j {X k : k i, j} for Gaussian case.

17 Examples of Label Dependent Sparsity (i) X stationary σ(i, j) = σ( i j ) spectral density f, 0 < ε f 1 ε < Ergodic ARMA processes satisfy (ii) T = S + K X = Y + Z Y, Z independent, S Y, K Z, S = [s(i j)] as in (i), K Hilbert-Schmidt: K 2 p (i, j) < (Z m 0, non stationary) i,j (i) s 2 (i) < i S = 0 Σ singular but correspond to functional data analysis A typical goal: Estimate Σ 1 for prediction.

18 Permutation Invariant Sparsity of Σ or Σ 1 a) Each row of Σ sparse or sparsely approximable e.g. If σ i = {σ i,j : 1 j p}, σ i 0 s for all i. b) Each row of Σ 1 sparsely approximable. a) roughly implies b) if λ min (Σ) δ > 0. A more general notion (El Karoui (2007) AS to appear). Roughly) The number of loops of length k in the undirected graph: i j iff σ ij 0 grows more slowly than p k/2 as k. The corresponding definition for Σ 1 is not equivalent in general.

19 Permutation Invariant Sparsity of Σ 1 Σ 1 corresponds to a graphical model. N (i) = { j : σ ij 0, j i } Sparsity means neighbourhood size p. Example: Gene networks Goal : Determine N (i) i = 1,..., p.

20 Construction of Estimates of Σ or Parameters of Σ which work well for T having sparse members Eigenstructure sparsity a) Sparse PCA d Aspremont et al., SIAM Review (2007), Hastie, Tibshirani, Zhou (2007). PCA with Lasso penalty on coordinates of eigenvectors Focus on computation, no asymptotics

21 Construction of Estimates of Σ or Parameters of Σ which work well for T having sparse members Eigenstructure sparsity continued b) Johnstone and Lu, JASA (2006). Sparse PCA for Σ corresponding to signals X = (X (t 1),..., X (t p)) T Represent in wavelet basis Use only k << p coefficients to represent data to whose covariance matrix PCA is applied Asymptotics for large p and examples

22 Construction of Estimates of Σ or Parameters of Σ which work well for T having sparse members Label Dependent Sparsity Banding Bickel and Levina (2004), Bernoulli Bickel and Levina (2007), Annals of Statistics (To appear) a) Banding Σ For any M m ij, B k (M) m ij 1( i j k) Σ 1 n (X i n X)(X i X) T Estimate: B k ( Σ), k = o(p). i=1

23 Construction of Estimates of Σ or Parameters of Σ which work well for T having sparse members Label Dependent Sparsity continued b) Banding Σ 1 Σ 1 = AA T, A Lower triangular Estimate: Σ 1 ÂkÂT k, where Âk is obtained by regression. Pourahmadi and Wu (2003), Biometrika. Huang,Liu,Pourahmadi (2007)

24 Construction of Estimates of Σ or Parameters of Σ which work well for T having sparse members Permutation Invariant Sparsity (a) Thresholding For any M = m ij, T t (M) m ij 1( m ij t) Estimate: T t ( Σ), t = o(1) El Karoui (2007) Bickel and Levina (2007).

25 Construction of Estimates of Σ or Parameters of Σ which work well for T having sparse members Permutation Invariant Sparsity continued b) SPICE Estimate of Σ 1 : Σ 1 sparse F (P, M) Trace(MP) log det(p) + λ P 1, where M 1 = i,j m ij { } a) ˆP (1) arg min F (P, ˆΣ) : P 0 Yuan, Li (2007), Biometrika. Fixed p asymptotics Banerjee et al. (2006), ICML computation { } b) ˆR (2) arg min F (P, ˆR) : R 0, r ii = 1, all i ˆP (2) ˆD ˆR ˆD, ˆD diag(ˆσ ii). ˆR ˆD 1 Σ ˆD 1 a),b): Rothman et al. (2007), Submitted to JRSS(B) Asymptotics

26 Construction of Estimates of Σ or Parameters of Σ which work well for T having sparse members Permutation Invariant Sparsity continued c) Methods for neighbourhood estimation Meinshausen and Buhlmann (2006), Zhao and Yu (2007): Regression methods + Lasso Wainwright (2006): Bounds Kalisch and Buhlmann (2007): Using the PC algorithm of Spertes et al. (2000) to roughly hierarchically test whether partial correlation coefficients are 0

27 Some Asymptotic Theory Matrix norms Operator norms M m ij p p p x r r x j r, x = (x 1,..., x p ) j=1 M (2,2) = λ 1/2 max(mm T ) p M (1,1) = max j m ij M (, ) = max i i=1 p m ij j=1 M max m ij i,j M 2 F i,j m 2 ij : Frobenius norm

28 For any operator norm, Properties AB A B Henceforth, M (2,2) M. If M p p is symmetric, { λ (M) M = Max, M [ ] 1/2. M (1,1) M (, ) M M F. Max λ (M) Min }.

29 Basic Property of Given A n, B n symmetric A n B n 0, suppose λ 1 (B n ) > λ 2 (B n ) > > λ k (B n ) > λ k+1 (B n ) and define λ j (A n ) analogously. Suppose λ j+1 (B n ) < λ j (B n ), 1 j k Dimension B n arbitrary, k, > 0 fixed. Then, a) λ j (A n ) λ j (B n ) = O( (j 1) A n B n ) b) If E ja respectively E jb is projection operator onto eigenspace corresponding to λ j, then E ja E jb = O( j A n B n )

30 B-L (2006) Main Result I Banded estimator : ˆΣ k,p (i, j) = ˆΣ p (i, j) 1( i j k) Let Theorem 1 U(ɛ 0, α, C) = { Σ : 0 < ε 0 λ min (Σ) λ max (Σ) 1/ε 0, max { σ ij : i j > k} Ck α for all k 0 }. j i If X is Gaussian and k n ( n 1 log p ) 1 2(α+1), then, uniformly on Σ U(ε 0, α, C), ˆΣ kn,p Σ p = O P ( (n 1 log p ) α 2(α+1) ) = ˆΣ 1 k n,p Σ 1 p The banded estimator and its inverse are consistent if log p n 0 Remark λ min ε 0 not needed if only convergence in to Σ is needed.

31 An Approximation Theorem (Bickel and Lindner (2007)) If m B k A δ, (B k banded of width 2k), A = M <, A 1 = m 1 <, K M m, There exists N(m, M, δ), B kn such that B kn A 1 1 c(m, M, δ)δ, M where c(m, M, δ) tends to c(k) as δ 0. B kn has a simply computable form.

32 Thresholding Theorem(Thresholding) (B-L Submitted to AS) p If U t (q, c 0 (p), M) = Σ : σ ii M 0, σ ij q c 0 (p), all i j=1 log p 0 q < 1, t n = M n, ) ( ) ) log p (1 q)/2 T tn (ˆΣ Σ = O p (c 0 (p) n, Note q = 0 c 0 (p) : non zero entries per row More subtle consistency results : N. El Karoui (2007).

33 SPICE (Rothman et al.) If S { (i, j) : σ ij 0 }, S s, Σ 1 m 1, Σ M and X is Gaussian, ˆP (1) Σ 2 F ( = O p (p + s) log p ) n ) ( ˆP (2) Σ 2 s log p = O p n

34 Choosing the Banding (or Thresholding or SPICE) Ideally want to minimize risk Parameter R(k) = E ˆΣ k Σ Estimate via a resampling scheme: Split the data into two samples of size n 1, n 2, N times at random Let ˆΣ (ν) 1, ˆΣ (ν) 2 be the two sample covariance matrices from the ν-th split. The risk can be estimated by ˆR(k) = 1 N (ˆΣ (ν) 1 N ) k ˆΣ (ν) 2 ν=1 We used n 1 = n/3, N = 50, and the L 1 matrix norm instead of L 2. Oracle properties for Frobenius norm (Bickel and Levina (2007), Annals of Statistics submitted).

35 Covariance of AR(1): Σ U Simulation Examples σ ij = ρ i j n = 100, p = 10, 100, 200, ρ = 0.1, 0.5, 0.9. Fractional Gaussian noise (FGN): long-range dependence, not in U σ ij = 1 2 [( i j + 1) 2H 2 i j 2H + ( i j 1) 2H] H [0.5, 1] is the Hurst parameter H = 0.5 is white noise; H = 1 is perfect dependence n = 100, p = 10, 100, 200, H = 0.5, 0.6, 0.7, 0.8, 0.9.

36 True and Estimated Risk for AR(1) 2 p = 10, ρ = p = 10, ρ = p = 10, ρ = True risk Estimated risk k k k p = 100, ρ = p = 100, ρ = p = 100, ρ = k k k p = 200, ρ = k p = 200, ρ = k p = 200, ρ = k

37 (1, 1) Loss and Choice of k for AR(1) Mean(SD) p ρ k 1 ˆk ˆΣˆk Loss ˆΣ k (0.5) 0.0(0.2) (0.8) 2.0(0.6) (0.7) 8.9(0.3) (0.4) 0.1(0.3) (0.7) 2.3(0.5) (4.5) 15.9(2.6) (0.4) 0.2(0.4) (0.7) 2.7(0.5) (4.5) 16.6(2.4) Table: AR(1): Oracle and estimated k and the corresponding loss values. k 1 = argmin ˆΣ k Σ (1,1). k ˆΣ

38 Operator Norm Results for Various Estimates and Situations p ρ Sample Band Band-P Thresh SPICE (0.01) 0.35(0.01) 0.35(0.01) 0.40(0.01) 0.35(0.01) (0.02) 0.64(0.02) 0.74(0.02) 0.80(0.02) 0.72(0.02) (0.05) 1.02(0.05) 1.02(0.05) 1.02(0.05) 1.01(0.05) (0.02) 0.45(0.01) 0.45(0.01) 0.50(0.01) 0.45(0.01) (0.03) 0.95(0.01) 2.04(0.00) 1.96(0.01) 1.52(0.01) (0.12) 4.45(0.09) 6.03(0.12) 6.18(0.12) 5.57(0.12) (0.02) 0.49(0.01) 0.49(0.01) 0.51(0.01) 0.50(0.01) (0.03) 1.00(0.01) 2.06(0.00) 2.04(0.00) 1.60(0.01) (0.16) 5.24(0.09) 9.76(0.16) 8.80(0.10) 7.50(0.10) Table: Operator Norm Loss, Averages and SEs over 100 replications. AR(1)

39 Operator Norm Results for Various Estimates and Situations p H Sample Band Band-P Thresh SPICE (0.01) 0.27(0.01) 0.27(0.01) 0.35(0.01) 0.28(0.01) (0.02) 0.88(0.02) 1.03(0.03) 0.88(0.04) 1.05(0.05) (0.04) 0.96(0.04) 0.96(0.04) 0.96(0.04) 1.13(0.05) (0.01) 0.38(0.01) 0.38(0.01) 0.46(0.01) 0.39(0.01) (0.05) 4.46(0.02) 5.37(0.00) 5.36(0.00) 5.30(0.02) (0.28) 8.03(0.28) 8.02(0.28) 8.02(0.28) 14.18(0.70) (0.02) 0.44(0.01) 0.44(0.01) 0.46(0.01) 0.45(0.01) (0.06) 6.46(0.01) 7.40(0.00) 7.40(0.00) 7.39(0.00) (0.43) 14.65(0.46) 14.34(0.43) 14.34(0.43) 30.16(0.60) Table: Operator Norm Loss, Averages and SEs over 100 replications. FGN

40 Example 1 continued EOF pattern #1 (Thresholding)

41 Example 1 continued EOF pattern #2 (Thresholding)

42 Some Important Directions Choice of k or other regularization parameters. Canonical correlation regularized estimation. Independent component analysis regularization. Estimation of parameters governing independence and conditional independence in graphical models.