## Current status linear regression

Current status linear regression Piet Groeneboom and Kim Hendrickx Delft University and University of Hasselt September 9, 2016 Figure: Kim Hendric...
Author: Jack Peters
Current status linear regression Piet Groeneboom and Kim Hendrickx Delft University and University of Hasselt

September 9, 2016

Figure: Kim Hendrickx

Current status regression Consider the regression model Yi = α + β 0 Xi + i , where Xi is a k-dimensional covariate and i an observation error with expectation zero. We can not observe Yi ! Instead, we observe (X1 , T1 , ∆1 ), . . . , (Xn , Tn , ∆n ), where Ti is an observation time and ∆i = 1{Yi ≤Ti } . The i are i.i.d. variables, independent of the Ti and Xi , with expectation zero; Ti and Xi are also taken to be independent. We want to estimate α and β. How to do this?

Maximum likelihood The log likelihood of the observations is n X

{∆i log FYi (Ti ) + (1 − ∆i ) log{1 − FYi (Ti )} ,

i=1

where  FYi (t) = P α + β 0 Xi + i ≤ t . If F is the distribution function of the i , we can write the log likelihood in the form n X 

∆i log F (Ti − α − β 0 Xi ) + (1 − ∆i ) log{1 − F (Ti − α − β 0 Xi ) .

i=1

We also have the extra condition that the expectation of i is zero, which amounts to: Z x dF (x) = 0.

Reduction to simpler model We drop the condition that the expectation of i is zero, and reduce the model to: Yi = β 0 Xi + i , where the i are i.i.d. with unknown expectation α. Then the log likelihood becomes: n X 

∆i log F (Ti − β 0 Xi ) + (1 − ∆i ) log{1 − F (Ti − β 0 Xi ) .

i=1

where F is the distribution function of i . ˆ ˆ We now first estimate R F and β by some estimates F and β, and then estimate α by x d Fˆ(x) (or a variation on this).

Example Example: Xi and Ti are uniform on [0, 2], that β = 1/2, α = Ei = 1/2, and suppose i has as density a rescaled version of the density 6x(1 − x) on [0, 1]. We rescale it to a density on [3/8, 5/8]: 6

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We can try to straightforwardly maximize w.r.t. F and β: n X i=1

{∆i log F (Ti − βXi ) + (1 − ∆i ) log{1 − F (Ti − βXi )}} .

Example (continued) Example for n = 1000: βˆ = 0.514828. For n = 10, 000: βˆ = 0.499999.

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Profile likelihood Profile likelihood: Pick a β, maximize for this β the log likelihood n X  ∆i log F (Ti − β 0 Xi ) + (1 − ∆i ) log{1 − F (Ti − β 0 Xi )} . i=1

over F , this gives Fˆβ via one step (convex minorant) algorithm. Now try to find an argmax for β over all Fˆβ so obtained. We use Brent’s algorithm for maximization, with Fˆβ as parameter. -0.02

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Figure: log likelihood for Fˆβ , as a function of β, n = 100.

Properties maximum likelihood estimator

What are the properties of the MLE of β so obtained? Unknown! √ Unknown whether the MLE of β is n-consistent Li and Zhang (1998) conjecture that the MLE will give a √ n-consistent but inefficient estimate. Murphy, van der Vaart, and Wellner (1999) prove that in a model in which one only has observations from a part of F0 where F0 stays away from 0 and 1 that the MLE gives a n1/3 -consistent estimate of β0 .

Interlude: binary choice model In econometrics, the binary choice model has been studied: special case of current status regression, where the Ti is degenerate at 0 and log likelihood of the form: n X 

∆i log F (β 0 Xi ) + (1 − ∆i ) log{1 − F (β 0 Xi )} .

i=1

The regression parameter β is only identifiable under an extra condition, for example kβk = 1. Example: Whether a person i says yes (∆i = 1) or no (∆i = 0) on a question in a questionnaire is predicted by his (socio-economic) covariate Xi . Accompanying theory is also still shrouded in mystery. Literature: Cosslett (1983, 2007), Klein and Spady (1993), Dominitz and Sherman (2005), lecture notes Bruce Hansen (2009).

Difficulties Why does estimating β seem so difficult? β is a parameter which appears as argument of a function F which √ we cannot estimate nonparametrically at rate n. Nevertheless, we expect that it is possible to estimate β at rate √ n. √ So we have to “bypass” the non- n estimable parameter F . Compare with proportional hazards model under current status: in this case the log likelihood is of the form n n o o  n X ∆i log 1 − exp −Λ(Ti )e βXi − (1 − ∆i )Λ(Ti )e βXi , i=1

where Λ is the baseline cumulative hazard function. Now β is not √ hiding inside a function F which is not- n estimable! Ordinary maximum likelihood is efficient (Huang (1996))!

Smoothing Consider restricting the estimates of F to a smooth plug-in estimator (Nadaraya Watson statistic): R δKh (t − βx − u + βy ) dPn (u, y , δ) , Fnh,β (t − βx) = R Kh (t − βx − u + βy ) dGn (u, y )

(1)

where, for a smooth symmetric kernel K : Kh (u) = h−1 K (u/h). Another way of writing Fn,β is to use ordinary sums: Pn j=1 ∆j Kh (t − βx − Tj + βXj ) Fnh,β (t − βx) = Pn . j=1 Kh (t − βx − Tj + βXj ) This type of estimate has also been considered in the econometrics literature. Note: Fn,β is not necessarily monotone, so need not be a distribution function!

Example plugin estimator

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Figure: Plug-in estimator, n = 1000 and βˆ = 0.49532.

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∂ log likelihood(Fnh,β ) = scores of plugin as function of β ∂β ∂ n 1 X {∆i − Fnh,β (Ti − βXi )} ∂β Fnh,β (Ti − βXi ) = n Fnh,β (Ti − βXi ){1 − Fnh,β (Ti − βXi )} i=1

Questions “Shrouded in mystery”: How should we choose the bandwidth h? Can we choose h  n−1/5 ? Can we use ordinary kernels? Cosslett (2007): we should have n−1/5 < h < n−1/8 (excluding h  n−1/5 and h  n−1/8 ). Klein and Spady (1993): n−1/6 < h < n−1/8 and one should take higher order (4th order) kernels. Lecture notes Bruce Hansen (2009): “It is unclear to me if these are merely technical sufficient conditions, or if there is a substantive difference with the semiparametric regression case.” Note: The Klein-Spady estimates of the finite-dimensional √ regression parameter have been proved to be n consistent and to achieve the information lower bound.

Efficiency plug-in estimators Theorem (Piet Groeneboom and Kim Hendrickx (2016)) Let Fnh,βˆn maximize the truncated log likelihood X

h ∆i logF (Ti − β 0 Xi )

F (Ti −β 0 Xi )∈[,1−]

+ (1 − ∆i ) log{1 − F (Ti − β 0 Xi )}

i

over all plug-in estimators (ordinary kernels). As n → ∞, and h  n−1/5 , then √

d

n(βˆn − β0 ) → N(0, I (β0 )−1 )

where I (β0 ) is the matrix with elements Z covar(Xi , Xj |T − β00 X = u) f0 (u)2 fT −β00 X (u) du. F (u){1 − F (u)} 0 0 F0 (u)∈[,1−]

Penalized estimates Murphy, van der Vaart, and Wellner (1999) consider the penalized maximum likelihood estimator, obtained by maximizing n X

{∆i log F (Ti − βXi ) + (1 − ∆i ) log{1 − F (Ti − βXi )}

i=1

Z − λn where

  1/λn = Op n2/5 ,

F 00 (u)2 du,

  λ2n = op n−1/2 ,

and observations are only on region where  < F0 (u) < 1 −  for some  > 0. √ Translated into bandwidth choice (hn  λn ), the penalty condition correspond to: n−1/5 ≤ h < n−1/8 .

Is smoothing really necessary? Consider the following estimator: βˆn is a value of β such that n X

 Xi ∆i − Fˆn,β (Ti − βXi ) = 0,

i=1

where Fˆn,β is the MLE of F0 based on the values Ti − βXi .

Theorem (Piet Groeneboom and Kim Hendrickx (2016)) D √  n βˆn − β0 −→ N(0, σ 2 ), where R

var(X |T − β0 X = u)F0 (u){1 − F0 (u)} fT −β0 X (u) du . R 2 var(X |T − β0 X = u)f0 (u) fT −β0 X (u) du √ Consequence: One can construct n-consistent estimates of β0 on the basis of the non-smoothed MLE’s Fˆn,β ’s. 2

σ =

Simple score function for MLE

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Figure: Simple score function for MLE, n=1000.

Maximum rank estimator of Arag´on and Quiroz (1995)

The regression parameter β is estimated by the maximizer βˆn of Γn (β) =

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∆i Ri (β),

i=1

where Ri (β) is the rank of Ti − βXi in T1 − βX1 , . . . , Tn − βXn . Similar to Han’s maximum rank correlation estimator (Han (1987)) in the econometric literature. √ Abrevaya (1999) proves that βˆn is n-consistent and asymptotically normal, using methods of Sherman (1993) in treating Han’s maximum rank correlation estimator.

Proof for maximum rank estimator Use Theorem 1 in Sherman (1993): kβˆn − β0 k = Op (n−1/2 ), ˆ if βn is the maximizer of Γn (β), with population equivalent Γ(β) and (a) there exists a neighborhood N of β0 and a constant k > 0 such that Γ(β) − Γ(β0 ) ≤ −kkβ − β0 k2 , for β ∈ N, and (b) uniformly over op (1) neighborhoods of β0 , Γn (β) − Γn (β0 )

 √  = Γ(β) − Γ(β0 ) + Op kβ − β0 k/ n + op kβ − β0 k2  + Op n−1 .

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(d) n = 1000 Figure: The log likelihood (red) and the criterion function Γn (blue) for sample sizes n = 50, 100, 500 and 1000. β0 = 1 is the true regression parameter. Ti , Xi and i are independent and standard normal.

Cumulative sum diagram for different β MLE Fˆn,β has jumps at a subset of {Ti − β 0 Xi , i=1,. . . ,n}. 1.0

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 P Fˆn,β (u) → P ∆i = 1|Ti − β 0 Xi = u Z = F0 (u + (β − β0 )0 x)fX |T −β 0 X (x|T − β 0 X = u) dx

Efficiency The efficient variance for estimating β is: Z var(X |T − β0 X = u)f0 (u)2 fT −β0 X (u) du. F0 (u){1 − F0 (u)} We can change βˆn to a value of β such that n X

 Xi φn (Ti − βXi ) ∆i − Fˆn,β (Ti − βXi ) = 0,

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where Xi φn (Ti − βXi ) estimates the efficient score, for example Xi φn (Ti − βXi ) = where fˆn,h (t) =

R

Xi fˆn,h (Ti − βXi ) . Fˆn,β (Ti − βXi ){1 − Fˆn,β (Ti − βXi )}

Kh (t − u) d Fˆn,β (u).

Theorem (Piet Groeneboom and Kim Hendrickx (2016)) Let βˆn be a value of β such that Z  xφn (t − β 0 x)) δ − Fˆn,β (t − β 0 x) dPn = 0, Fˆn,β ∈[,1−]

where: φn (t − βx) = and fˆn,h (t) =

R

fˆn,h (t − β 0 x) , Fˆn,β (t − β 0 x){1 − Fˆn,β (t − β 0 x)}

Kh (t − u) d Fˆn,β (u). Then D √  n βˆn − β0 −→ N(0, I (β0 )−1 ),

Z I (β0 ) = F0 ∈[,1−]

var(X |T − β00 X = u)f0 (u)2 fT −β00 X (u) du . F0 (u){1 − F0 (u)}

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Performance of estimates of β

n 100 500 1000 5000 10000 20000

Plug-in, β via argmax mean(βˆn ) nvar (βˆn ) 0.499562 0.245172 0.498857 0.191857 0.499502 0.192223 0.500314 0.181421 0.500120 0.172043 0.500096 0.174197

MLE, β via score=0 mean(βˆn ) nvar (βˆn ) 0.502964 0.362099 0.500167 0.212293 0.500178 0.194511 0.500058 0.176652 0.499989 0.174698 0.500001 0.169884

MLE, β via argmax mean(βˆn ) nvar (βˆn ) 0.487057 0.405507 0.498111 0.306398 0.499528 0.296097 0.500162 0.227073 0.500159 0.233758 0.500050 0.236070

Table: h = 0.5n−1/5 and N = 1000. (I (β0 )−1 = 0.158699,  = 0.001)

Efficiency

Conclusion: We can construct an asymptoticallly efficient estimate with the non-smoothed MLE, using an estimate of the density based on the MLE. This uses isotonic estimators and reordering properties.

Bibliography I Jason Abrevaya. Rank regression for current-status data: asymptotic normality. Statist. Probab. Lett., 43(3):275–287, 1999. ISSN 0167-7152. URL http://dx.doi.org/10.1016/S0167-7152(98)00267-3. Jorge Arag´on and Adolfo J. Quiroz. Rank regression for current status data. Statist. Probab. Lett., 24(3):251–256, 1995. ISSN 0167-7152. URL http://dx.doi.org/10.1016/0167-7152(94)00180-G. Stephen R. Cosslett. Distribution-free maximum likelihood estimator of the binary choice model. Econometrica, 51(3): 765–782, 1983. ISSN 0012-9682. URL http://dx.doi.org/10.2307/1912157. Stephen R. Cosslett. Efficient estimation of semiparametric models by smoothed maximum likelihood. Internat. Econom. Rev., 48 (4):1245–1272, 2007. ISSN 0020-6598. URL http://dx.doi.org/10.1111/j.1468-2354.2007.00461.x.

Bibliography II Jeff Dominitz and Robert P. Sherman. Some convergence theory for iterative estimation procedures with an application to semiparametric estimation. Econometric Theory, 21(4):838–863, 2005. ISSN 0266-4666. URL http://dx.doi.org/10.1017/S0266466605050425. Aaron K Han. Non-parametric analysis of a generalized regression model: the maximum rank correlation estimator. Journal of Econometrics, 35(2-3):303–316, 1987. Jian Huang. Efficient estimation for the proportional hazards model with interval censoring. Ann. Statist., 24(2):540–568, 1996. ISSN 0090-5364. URL http://dx.doi.org/10.1214/aos/1032894452. Roger W. Klein and Richard H. Spady. An efficient semiparametric estimator for binary response models. Econometrica, 61(2): 387–421, 1993. ISSN 0012-9682. URL http://dx.doi.org/10.2307/2951556.

Bibliography III

Gang Li and Cun-Hui Zhang. Linear regression with interval censored data. Ann. Statist., 26(4):1306–1327, 1998. ISSN 0090-5364. URL http://dx.doi.org/10.1214/aos/1024691244. S. A. Murphy, A. W. van der Vaart, and J. A. Wellner. Current status regression. Math. Methods Statist., 8(3):407–425, 1999. ISSN 1066-5307. Robert P. Sherman. The limiting distribution of the maximum rank correlation estimator. Econometrica, 61(1):123–137, 1993. ISSN 0012-9682. URL http://dx.doi.org/10.2307/2951780.