y Manajemen | Fakultas Ekonomi Universitas Maritim Raja Ali Haji 073500103288619331

Frühwirth-Schnatter, Tüchler, and Otter: Bayesian Analysis of the Heterogeneity Model 3 has been applied successfully in marketing science in the pio- neering work of Allenby et al. 1998 and Lenk and DeSarbo 2000. Although Algorithm 1 happened to work Žne in these applications, it is necessary to discuss potential pitfalls of this algorithm and to consider alternative methods of MCMC esti- mation of model 1. A Žrst potential drawback of Algorithm 1—either con- strained or unconstrained—is that in case that correlations are high between quantities in the different subblocks, especially within step iii, the resulting sampler will be slowly mixing. This is a well-discussed issue for the normal linear mixed model Gelfand, Sahu, and Carlin 1995 which is a special case of the heterogeneity model with K D 1. We demonstrate in this article that similar problems are to be expected for the general model where K 1 and the case study in Section 4.3 highlights the rather dramatic consequences for practical work. Algorithm 1 turns out to be sensitive to the parameterization used for the mean structure of model 1 and might exhibit slow conver- gence in cases where random effects with low variance are centered around the group means and in cases where random effects with high variance are centered around 0. As an alternative to Algorithm 1, we exploit a common way of overcoming mixing problems caused by correlation, namely “blocking” i.e., updating parameters jointly and “collapsing” i.e., sampling parameters from partially marginalized distribu- tions see Chen, Shao, and Ibrahim 2000; Liu, Wong, and Kong 1994. Such blocked and collapsed samplers have been derived for a normal linear mixed model independently by Chib and Carlin 1999 and Frühwirth-Schnatter and Otter 1999; in this article, we extend them to the mixture of random-effects mod- els 1 with K 1. To our knowledge, such a collapsed sam- pler has not yet been considered for the general heterogeneity model. This alternative sampler will turn out to be robust to the parameterization underlying the mean structure. The resulting sampler is only partly marginalized; in the sense that all vari- ance parameters are still sampled from their full conditionaldis- tributions. Marginalizingthis part of the sampling scheme is not possible without leaving the convenient Gibbs sampling frame- work. However, by introducing a Metropolis–Hastings step into the sampler, we are able to investigate the merits of “collaps- ing” this sampling step. Interestingly, this fully marginalized MCMC scheme in the sense that no sampling step conditions on the random effects does not provide a noticeable improve- ment over the partly marginalized sampler. The second topic of this article concerns the identiŽability problem inherent in model 1. Because model 1 includes the latent, discrete structure S N , the unconstrained posterior typically is multimodal. The impact of this type of unidenti- Žability on MCMC estimation of mixture models has been dis- cussed by Stephens 2000, Celeux, Hurn, and Robert 2000, and Frühwirth-Schnatter 2001a. In this article we add some more aspects on this identiŽability problem and illustrate for simulated data that a standard constraint as the one applied in step ii-1 of the constrained Algorithm 1 does not neces- sarily restrict the posterior simulations to a unique modal re- gion. Choosing of a suitable identiŽability constraint appears particularly difŽcult in the multivariate setting considered here. We will demonstrate within a case study from metric conjoint analysis how to explore MCMC simulations from the uncon- strained posterior to obtain an identiŽability constraint that is able to separate the posterior modes. The rest of the article is organized as follows. In Section 2 we discuss, starting from Algorithm 1, various MCMC sam- plers obtained from Algorithm 1 by grouping and collapsing. In Section 3 we explore the unidentiŽability problem. We present an empirical illustration involving metric conjoint analysis in Section 4, and conclude the article with some Žnal remarks in Section 5. 2. PARAMETERIZATION, GROUPING, AND COLLAPSING 2.1 Parameterization and Sampling From the Full Conditionals A straightforward method for MCMC estimation of the gen- eral heterogeneity model 1 is sampling from the full condi- tionals Algorithm 1 in Sec. 1. The problem with sampling from the full conditionals is that in cases where correlations are high between quantities in different subblocks, the resulting sampler will be slowly mixing. The possible inuence of the parameterization on the efŽciency of the MCMC sampler, espe- cially within step iii-1, is a well-discussed issue for the normal linear mixed model Gelfand et al. 1995. Because this model is the special case of 1 where K D 1, we expect a similar inu- ence of the parameterization for the more general model, which, conditional on S N , could be regarded as the combination of K normal linear mixed models with the random effects of each group independent from each other. 2.1.1 Centered and Noncentered Parameterization. The mean structure of model 1 can be parameterized in two com- mon ways. If X 1 i and X 2 i do not have any common column, then ® represents parameters that are Žxed for all individu- als, whereas ¯ i are the random heterogeneous effects cen- tered around ¯ G k , the population mean in class k: E.¯ i jS i D k; Á D ¯ G k . The expectation of the marginal distribution of y i , where both the random effects and the latent indicators are in- tegrated out, is given by

E. y

i jÁ D X 1 i ® C X 2 i ® ¯ ; ® ¯ D K X k D1 ¯ G k ´ k : 4 Conditionalon S i , centering is around the group-speciŽc means, whereas marginally where the indicators are integrated out, the random effects are centered around the population mean, E.¯ i jÁ D ® ¯ . Therefore, this parameterization seems to be rather close to the idea of hierarchical centering introduced by Gelfand et al. 1995, and thus we call it centered parameteriza- tion. This is the parameterization used by Allenby et al. 1998 and Lenk and DeSarbo 2000. Alternatively, Verbeke and Lesaffre 1996 designed mo- del 1 in such a way that X 1 i and X 2 i may have common columns. This implies the additional constraint that marginally where the indicators are integrated out the random effects are deviations from the Žxed effects with mean 0, E.¯ i jÁ D P K k D1 ¯ G k ´ k D 0. The Žrst two moments of the marginal dis- tribution of y i , where both the random effects and the latent Downloaded by [Universitas Maritim Raja Ali Haji] at 00:19 13 January 2016 4 Journal of Business Economic Statistics, January 2004