Genome-wide linkage analysis with SNPs

A genome-wide linkage scan of the count variables available in 1519 individuals in 710 families using 5744 SNPs (Illumina panel) was performed via MERLIN-REGRESS (version 1.0.1) [3], with the square root transformation of the counts treated as quantitative traits. In this regression-based method [4], we used the sample estimates of the transformed counts as population means and variances, and estimated heritability using the variance-components (VC) option in MERLIN (version 1.0.1) [3], with values of 39% and 42% for transformed swollen and tender joint counts, respectively. Because genetic maps were not available for the SNPs, we assumed that 1 Mb is equivalent to 1 cM.

Directing our attention to chromosome 6, which has been previously reported to show a significant region of linkage (harboring the HLA-DRB1 locus) for RA, we found two regions of modest signal for each trait (see Fig. 1). One region detected for both count variables spanned 26 to 33 Mb. A second region for the tender count variable spanned 114 to 117 Mb, and a differing secondary region for the swollen count variable spanned 155 to 158 Mb. Several other regions of interest were also detected on other chromosomes (Fig. 1). Adjusting this multipoint analysis for linkage disequilibrium using MERLIN produced minor changes (Fig. 1).

A bivariate counting process model for genetic association of RA severity

Methods for analyzing data on events observed over time have been of considerable interest in recent years. A counting process framework [5, 6] models count data, collected at fixed points of ascertainment, by taking duration of disease into consideration (as is the case in a cross-sectional design, provided that the time of disease onset is available). Under the assumption that the numbers of tender and swollen joints do not decrease through time, we have a recurrent event process.

Let {N _ij (t), t ≥ 0} represent the underlying counting process in which N _ij (t) denotes the number of events experienced by the j^th process of the i^th individual over the continuous time interval (0, t], where i = 1,..., n. Note that j = 1, 2 represents the tender and swollen joint counting processes, respectively. Because it is reasonable to specify a model that depends only on the current covariate values rather than on the entire history process in our setting, we formulated a non-homogeneous Poisson process. To handle any substantial inter-individual variation in the model, we introduced a positive individual-specific random effect u _i that is shared for both processes of the i^th subject. Conditional on u _i , the counts of the i^th subject are independent. To be more specific, we assumed that conditional on u _i , the counting process {N _ij (t), t ≥ 0} is a non-homogeneous Poisson process with intensity and mean functions represented as

λ _ij (t|u _i ) = u _i λ_0j(t) exp{x _ij ^Tβ _j } (1)

and

Λ _ij (t|u _i ) = u _i Λ_0j(t) exp{x _ij ^Tβ _j }, (2)

respectively. The time line is defined as the time since onset of RA. The form of the intensity function is a relative risk or Cox model, with the baseline intensity function for the j^th process λ_0j(t) common among all subjects. The covariate vector x _ij , specific to the i^th subject, may include candidate polymorphisms (such as DRB1), single SNPs selected through genome-wide linkage scans, or a vector of SNPs chosen to capture variation in a candidate gene or candidate region (such as PTPN22); it may also include characteristics such as sex and smoking history, or any combination of reasonable interactions. The corresponding regression parameter vector β _j describes the association of genotype with phenotype for the j^th process. Furthermore, each individual has their own frailty u _i acting multiplicatively on the intensity function.

Unless the variance of the underlying mixing distribution is exceptionally large, the gamma random effect provides a robust approach for modeling mixed-Poisson processes [7]. Assuming that u₁,..., u _n are independent and identically distributed random variables arising from a gamma distribution with mean of 1 and variance of φ, we obtain the expectation E[N _ij (t)] = E (E[N _ij (t)|u _i ]) = Λ_0j(t) exp{x _ij ^Tβ _j }, which represents the mean number of counts for the j^th process of the i^th individual over the interval (0, t]. The parameter φ is a measure of heterogeneity between individuals that may not have been sufficiently accounted for by the Poisson model alone. Larger values of φ imply extra-Poisson variation in the model.

We now need to construct a likelihood function based on our bivariate mixed-Poisson process assumptions. Suppose the i^th individual is ascertained at time τ _i , which is relative to the time of onset. Conditional on the random effect u _i , the distribution of the counts P(N _ij (τ _i ) = n _ij |u _i ) has a Poisson form. Moreover, under conditional independence, the bivariate distribution of the counts can be computed as P(n_i1, n_i2) = ∫ P(n_i1, n_i2|u _i ) dG(u _i ) = ∫ P(n_i1|u _i ) P(n_i2|u _i ) g(u _i ) du _i , which has a convenient closed form expression due to the gamma-Poisson mixture. Thus the log-likelihood log L(θ) = Σ log P(n_i1, n_i2) is maximized with respect to θ, which consists of parameters β₁, β₂, φ, and parameters of the baseline intensity functions. We applied a Newton-Raphson technique, but any non-linear maximization algorithm may be used for this purpose. Furthermore, when maximizing the log-likelihood under the counting process framework, it is most convenient to assume a parametric form such as a Weibull model for the baseline intensity functions, although semi-parametric assumptions using piecewise constant models are also reasonable alternatives.

In the case of a univariate counting process model, there is no joint distribution formulated among the counts for each individual. Rather, the log-likelihood is simply L(θ) = ΣΣ log P(n _ij ), where P(n _ij ) = ∫ P(n _ij |u _i ) dG(u _i ) = ∫ P(n _ij |u _i ) g(u _i ) du _i .

Association analysis of candidate gene DRB1

We applied our bivariate mixed-counting process model for the severity phenotypes under various covariate models, yielding estimates and standard errors of β₁, β₂, and log φ (Table 1). The asymptotic confidence interval for log φ indicated a significant amount of extra-Poisson variation in the counts. That is, the number of tender joints varied considerably between patients, as did the number of swollen joints. The heterogeneity appeared to decrease as more covariates were added to the model, and any remaining variation in the model was captured under this random effects formulation. The log-likelihood increased dramatically as the covariates sex, smoking history, and DRB1 were added to the model. The likelihood ratio (LR) test for the joint contribution of sex to the bivariate model (2 degrees of freedom) provided a p-value less than 1 × 10^-6, and the LR test for the joint contribution of DRB1 provided a p-value less than 0.0001. The Wald test detected a significant sex effect for the tender, but not for the swollen joint process. These sex differences were evident in plots of the estimated mean number of counts (Fig. 2). The LR test for the equality of DRB1 between the two counting processes in the bivariate model (1 degree of freedom) yielded a p-value less than 1 × 10^-5.

				Covariate associations
Model	log L	Heterogeneity log φ (SE)	Outcome	Sex β (SE)	Smoking history β (SE)	DRB1 β (SE)
No covariates	-10052.06	-0.358 (0.04)	Tender
			Swollen
With sex	-10035.23	-0.360 (0.04)	Tender	-0.185 (0.05)
			Swollen	-0.010 (0.05)
With sex,	-10031.06	-0.365 (0.04)	Tender	-0.216 (0.06)	0.139 (0.04)
smoking history			Swollen	-0.035 (0.05)	0.110 (0.04)
With sex,	-10020.33	-0.366 (0.04)	Tender	-0.216 (0.06)	0.139 (0.05)	-0.009 (0.03)
smoking history, and DRB1			Swollen	-0.038 (0.06)	0.107 (0.05)	0.071 (0.03)

Relative efficiency computations (not shown), obtained by taking the ratio of the variances, suggested a gain in information for the bivariate versus univariate counting process model. For the tender and swollen outcomes, the relative variance ranged from 1.122 to 1.294 and 1.005 to 1.015, respectively, under various covariates.

Association analysis of the 14 PTPN22 SNPs on chromosome 1 and 404 illumina SNPs on chromosome 6

To evaluate the genetic association of candidate gene PTPN22 and SNPs on chromosome 6, we performed various LR tests under both univariate and bivariate mixed-counting process models. Under the univariate model we tested

(i) H _o : β_{tender, SNP}= 0 and (ii) H _o : β_{swollen, SNP}= 0,

and under the bivariate model we tested

(i) H _o : β_{tender, SNP}= 0, β_{swollen, SNP}= 0 and (ii) H _o : β_{tender, SNP}= β_{swollen, SNP}.

Note that, along with sex and smoking history, DRB1 was included in the model since it is significant based on the results of the association analysis above, and also because it has been previously reported to show strong linkage [1] and association with RA. Figure 3 consists of plots of the p-values for each SNP. The dashed line indicates a region-wide Bonferroni significance criterion, computed as the ratio of a selected significant p-value (0.05) over the number of tests performed.