Abstract
We show that in a large class of proportional hazard competing risks models, the distribution of the bivariate frailty among survivors converges to a limiting distribution. This generalizes the result of Abbring and van den Berg (Biometrika 94(1):87–99, 2007), who show that in a single spell duration model, the frailty distribution converges to the gamma distribution. The resulting limiting distribution has an interpretation in terms of partition of a gamma distribution, and allows for both positive and negative correlation between the survival variables. This result provides a natural and flexible specification for the frailty in competing risks models.
Similar content being viewed by others
Notes
The conventional approach is to use point mass distributions see e.g., Deng et al. (2000).
The second condition is to ensure that the probability of surviving infinitely is zero, that is, there is no defectors.
Because of the cointegration relation, it is also equivalent to check the potential convergence of \((\Lambda _1(t)U, \Lambda _2(t))|T_1>t, T_2>t\).
This is important since we have (informally) \(\mathbb {1}_{X_t<Y_t} \approx \mathbb {1}_{T_1<T_2}\), namely, the order is (asymptotically) preserved under the time change.
A function F is \(RV_0\) with index \(\alpha >0\) if and only if:
$$\begin{aligned} \displaystyle \lim _{y \rightarrow 0} \frac{F(y)}{y^{\alpha }L(y)}=1, \end{aligned}$$where L is a slowly varying function at zero, that is, \(\lim _{y \rightarrow 0}\frac{L(ay)}{L(y)}=1\) for any \(a>0\).
Indeed, the primary aim of the paper is to argue that the distribution \(dF(u,v) \propto e^{-a_1u-a_2v} \mathrm {d} \nu (u,v)\) is a natural and convenient choice for the initial distribution.
See however the working paper version of Gordon (2002), available at http://www.polmeth.wustl.edu/files/polmeth/gordo00.pdf, for another model with negatively correlated frailties. Gordon assumes \(U=1/V\) with U gamma distributed.
Such as a common heterogeneity term for the two risks.
Such as the mixture of point masses.
References
Abbring J, van den Berg G (2003) The identifiability of the mixed proportional hazards competing risks model. J R Stat Soc B 65(3):701–710
Abbring J, van den Berg G (2007) The unobserved heterogeneity distribution in duration analysis. Biometrika 94(1):87–99
Baker M, Melino A (2000) Duration dependence and nonparametric heterogeneity: a Monte–Carlo study. J Econ 96(2):357–393
Clayton D (1978) A model for association in bivariate life tables and its application in epidemiological studies of familial tendency in chronic disease incidence. Biometrika 65(1):141–151
Deng Y, Quigley J, Order R (2000) Mortgage terminations, heterogeneity and the exercise of mortgage options. Econometrica 68(2):275–307
Feller W (2008) An introduction to probability theory and its applications, vol 2, 2nd edn. Wiley, Hoboken
Gordon SC (2002) Stochastic dependence in competing risks. Am J Polit Sci 46:200–217
Hanagal DD (2009) Weibull extension of bivariate exponential regression model with different frailty distributions. Stat Pap 50(1):29–49
Heckman J, Honoré B (1989) The identifiability of the competing risks model. Biometrika 76(2):325–330
Horny G (2009) Inference in mixed proportional hazard models with k random effects. Stat Pap 50(3):481–499
Kayid M, Izadkhah S, Zuo MJ (2015) Some results on the relative ordering of two frailty models. Stat Pap 58:1–15
Lancaster T (1979) Econometric methods for the duration of unemployment. Econometrica 47(4):939–956
Ledford AW, Tawn JA (1996) Statistics for near independence in multivariate extreme values. Biometrika 83(1):169–187
Lukacs E (1955) A characterization of the gamma distribution. Ann Math Stat 26:319–324
Omey E, Willekens E (1989) Abelian and Tauberian theorems for the laplace transform of functions in several variables. J Multivar Anal 30(2):292–306
Xue X, Brookmeyer R (1996) Bivariate frailty model for the analysis of multivariate survival time. Lifetime Data Anal. 2(3):277–289
Yashin A, Vaupel J, Iachine I (1995) Correlated individual frailty: an advantageous approach to survival analysis of bivariate data. Math Popul Stud 5(2):145–159
Acknowledgements
I thank C. Gouriéroux, A. Guillou, G. Stupfler, C. Genest, as well as two referees for insightful comments. All remaining errors are mine.
Author information
Authors and Affiliations
Corresponding author
Appendices
Appendix 1
Proofs
Proof of Property 2
which converges to a positive number \(\ell '\) if the distribution \(\Lambda (t)(U, V)| T_1>t, T_2>t\). Thus \(\mathbb {P}[T_1<T_2|\min (T_1, T_2)=t]=\frac{h_1(t)}{h_1(t)+h_2(t)}\) converges to \(\frac{\ell '}{1+\ell '}\).
Proof of Theorem 1
The convergence of the distribution of \((\Lambda (t)U,\Lambda (t)V)|T_1>t, T_2>t\), or equivalently that of \((\Lambda _1(t)U,\Lambda _2(t)V)| T_1>t, T_2>t\), is equivalent to the convergence of the Laplace-Stieltjes transform of the cdf of the latter:
Thus its convergence implies the pointwise convergence of: \(\frac{F(\frac{u}{\Lambda _1(t)}, \frac{v}{\Lambda _2(t)})}{\mathbb {E}\big [e^{-\Lambda _1(t)U-\Lambda _2(t)V}\big ]}\) to a measure k(u, v), say. In particular, by taking \(u=v=1\) we also get: \(\displaystyle \frac{F(\frac{u}{\Lambda _1(t)}, \frac{v}{\Lambda _2(t)})}{F(\frac{1}{\Lambda _1(t)}, \frac{1}{\Lambda _2(t)})} \rightarrow \frac{k(u, v)}{k(1, 1)}\), or equivalently (by the monotonic property of F): \(\displaystyle \frac{F(\frac{u}{\Lambda (t)}, \frac{v}{\Lambda (t)})}{F(\frac{1}{\Lambda (t)}, \frac{1}{\Lambda (t)})} \rightarrow \frac{k(u/a_1, v/a_2)}{k(1/a_1, 1/a_2)}\). Thus F is regularly varying at zero.
Conversely, if F is regularly varying at zero, then, under some regularity conditions, the extended continuity theorem (Feller 2008, Theorem 2, Chapter XIII.1) applies and the previous steps can be reversed. \(\square \)
Proof of Property 9
(1) By Feller (2008), Chapter XIII.5, the regular variation at zero of \(a_1 U+a_2 V\) is equivalent to the regular variation of its Laplace transform at infinity. When t goes to infinity, we have:
by the homogeneity of \(\nu \). Thus \(a_1 U+a_2 V\) is \(RV_0(\alpha )\).
Proof of Property 5
If (U, V) follows (6), then the pdf of \(\Lambda (t)(U, V)\) given \(T_1>t, T_2>t\) is
Since \(\Lambda _1(t) \sim a_1 \Lambda (t)\), \(\Lambda _2(t) \sim a_2 \Lambda (t)\), \(l(\frac{u}{\Lambda (t)}, \frac{v}{\Lambda (t)}) \sim l(\frac{1}{\Lambda (t)}, \frac{1}{\Lambda (t)})\), and \(\mu \) is homogeneous, the distribution (19) converges to: \(f_{\infty }(u, v) \propto e^{-a_1u-a_2v}\mu (u, v)\). Therefore, (U, V) is \(BRV_0(\nu )\) by Theorem 1.
Conversely, if the cdf F is regularly varying at zero with a limit measure \(\nu \), then, under some regularity conditions (see e.g., Omey and Willekens 1989), the pdf f is also regularly varying, with:
Let us define \(l(x, y)=\frac{f(x, y)}{\mu (x, y)}\); then by Eq. (20), we can check that l is slowly varying at zero. \(\square \)
Rights and permissions
About this article
Cite this article
Lu, Y. The distribution of unobserved heterogeneity in competing risks models. Stat Papers 61, 681–696 (2020). https://doi.org/10.1007/s00362-017-0956-y
Received:
Revised:
Published:
Issue Date:
DOI: https://doi.org/10.1007/s00362-017-0956-y