Abstract
In this article, we propose a new joint modeling approach for the analysis of longitudinal data with informative observation times and a dependent terminal event. We specify a semiparametric mixed effects model for the longitudinal process, a proportional rate frailty model for the observation process, and a proportional hazards frailty model for the terminal event. The association among the three related processes is modeled via two latent variables. Estimating equation approaches are developed for parameter estimation, and the asymptotic properties of the proposed estimators are established. The finite sample performance of the proposed estimators is examined through simulation studies, and an application to a medical cost study of chronic heart failure patients is illustrated.
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Acknowledgements
The authors would like to thank the Editor-in-Chief, Professor Mei-Cheng Wang, an Associate Editor and two referees for their constructive and insightful comments and suggestions that greatly improved the article. This research was supported by GRF 14305014 from the Research Grant Council of the Hong Kong Special Administration Region, Direct Grants from the Chinese University of Hong Kong, the National Natural Science Foundation of China (Grant Nos. 11771431 and 11690015) and Key Laboratory of RCSDS, CAS (No. 2008DP173182), and the Fundamental Research Funds for the Central Universities (Grant No. 20205170465).
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Appendices
Appendix : Proof of Asymptotic Results
In order to study the asymptotic properties of the proposed estimators, we need the following regularity conditions:
-
(C1)
\(\{Y_i(\cdot ), N_i^R(\cdot ), N_i^D(\cdot ), T_i, X_i(\cdot ),\ i=1,\ldots , n\}\) are independent and identically distributed.
-
(C2)
\(E\{N_i^R(\tau )\}\) is bounded, and \(\Pr (T_i \ge \tau )>0\).
-
(C3)
\(X_i(t)\) is almost surely of bounded variation on \([0, \tau ]\).
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(C4)
There exist a compact set \(\mathcal {B}\) of \(\xi _0\) such that for \(\xi \in \mathcal {B}\), \(\Gamma (\xi )\) is nonsingular, where \(\Gamma (\xi )\) is the limit of \(-\partial \tilde{U}(\xi )/\partial \xi ^T\) with \(\tilde{U}(\xi )\) defined in (A.3).
For notational simplicity, henceforth we omit superscript t in the covariate \(X_i(t).\) Let \(\tilde{\Lambda }_0^D(t)\) and \(\tilde{\Lambda }_0^R(t)\) denote the solutions to the following estimating equations
and
Proof of Theorem 1
Define
where \(S_0(t; \alpha )=\frac{1}{n}\sum _{i=1}^n\Delta _i(t)\exp \{\alpha ^TX_i\}.\) By the definition of \(\tilde{\Lambda }_0^D(t),\) it can be checked that
which is a linear Volterra integral equation, and the solution is
where \(\tilde{P}(t)=\prod _{s\le t} \{1- d \tilde{\Psi }_1(s)\}\) is the product-integral of \(\tilde{\Psi }_1(s)\) over [0, t]. Using the uniform convergence of \(\tilde{\Lambda }_0^D(t),\) the uniform strong law of large numbers [12] and Lemma A.1 of Lin and Ying [8], we get that uniformly in \(t \in [0, \tau ],\)
where
and P(t) and \(s_0(t;\alpha )\) are the limits of \(\tilde{P}(t)\) and \(S_0(t;\alpha )\), respectively. Let
and \(\Psi _2(t)\) be the limit of \(\tilde{\Psi }_2(t).\) Note that
where \(\psi _i(t;\Lambda _0^D)=\psi _i(t;\alpha _0, \theta _0, \Lambda _0^D).\) It then follows from (A.1) that
where
Let \(\omega _{2i}(t; \xi )\) be defined as \(\omega _{2i}(t)\) with \(\psi _i(t)\) replaced by \(\psi _i(t;\alpha , \theta , \tilde{\Lambda }_0^D),\) and
where
and
with \(B_i(t; \tilde{\Lambda }_0^D)=B_i(t; \alpha , \theta , \tilde{\Lambda }_0^D),\) and
Note that
where \(\psi _i(t; \tilde{\Lambda }_0^D)=\psi _i(t;\alpha _0, \theta _0, \tilde{\Lambda }_0^D).\) Let
and \(\bar{x}(t;\alpha _0)\) and \(H_1(t)\) are the limits of \(\bar{X}(t; \alpha _0)\) and \(\tilde{H}_1(t),\) respectively. Similarly to (A.2), we have
where
Likewise,
where
and \(\bar{x}(t;\beta _0)\) and \(H_2(t)\) are the limits of \(\bar{X}(t;\beta _0)\) and \(\tilde{H}_2(t)\) with
Let
and q(t) be the limit of \(\tilde{Q}(t).\) Note that
As in the proof of (A.4), the second term on the right-hand side of (A.6) equals
where \(H_3(t)\) and \(H_4(t)\) are the limits of \(\tilde{H}_3(t)\) and \(\tilde{H}_4(t)\) with
and
Let \(\mathrm{{d}}\Phi (t)=E\big \{\omega _{2i}^*(t) \mathrm{{d}}N_i^D(t)\big \},\)
and \(H_5(t)\) and \(H_6(t)\) be the limits of \(\tilde{H}_5(t)\) and \(\tilde{H}_6(t),\) respectively. In a similar manner, the third term on the right-hand side of (A.6) is
Thus, it follows from (A.6)–(A.8) that
where
Let \(B_i(t)= B_i(t; \alpha _0, \theta _0, \Lambda _0^D),\) and
Note that
where
and
For \(R_1\) of (A.10), using the Lemma A.1 of Lin and Ying [8], we have
where \(\bar{b}(t)\) is the limit of \(\bar{B}(t).\) For \(R_2\), we get
It then follows from (A.1) and (A.2) that
where \(G_1(t)\) and \(G_2(t)\) are the limits of \(\tilde{G}_1(t)\) and \(\tilde{G}_2(t),\) respectively, with
By the Lemma A.1 of Lin and Ying [8], it can be shown that \(R_3=o_p(n^{1/2}).\) Thus, we have
where
Let \(\vartheta _i=(\vartheta _{1i}^T,\vartheta _{2i}^T,\vartheta _{3i},\vartheta _{4i}^T)^T,\) and \(\Gamma =\Gamma (\xi _0)\) defined in condition (C4). Then it follows from (A.4), (A.5), (A.9), (A.11), and the Taylor expansion that
By the multivariate central limit theorem, \(n^{1/2}(\hat{\xi }-\xi _0)\) is asymptotically normal with mean zero and covariance matrix \(\Gamma ^{-1}\Sigma (\Gamma ^T)^{-1},\) where \(\Sigma =E\{\vartheta _i \vartheta _i^T\}\).
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Qu, L., Sun, L. & Song, X. A Joint Modeling Approach for Longitudinal Data with Informative Observation Times and a Terminal Event. Stat Biosci 10, 609–633 (2018). https://doi.org/10.1007/s12561-018-9221-8
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DOI: https://doi.org/10.1007/s12561-018-9221-8