Risk Alleles for Systemic Lupus Erythematosus in a Large Case-Control Collection and Associations with Clinical Subphenotypes

doi:10.1371/journal.pgen.1001311

PLoS Genet. 2011 February; 7(2): e1001311.

Published online 2011 February 17. doi: 10.1371/journal.pgen.1001311

PMCID: PMC3040652

Copyright This is an open-access article distributed under the terms of the Creative Commons Public Domain declaration which stipulates that, once placed in the public domain, this work may be freely reproduced, distributed, transmitted, modified, built upon, or otherwise used by anyone for any lawful purpose.

Risk Alleles for Systemic Lupus Erythematosus in a Large Case-Control Collection and Associations with Clinical Subphenotypes

Kimberly E. Taylor,¹ Sharon A. Chung,¹ Robert R. Graham,² Ward A. Ortmann,² Annette T. Lee,³ Carl D. Langefeld,⁴ Chaim O. Jacob,⁵ M. Ilyas Kamboh,⁶ Marta E. Alarcón-Riquelme,^7,⁸ Betty P. Tsao,⁹ Kathy L. Moser,⁷ Patrick M. Gaffney,⁷ John B. Harley,^10,¹¹ Michelle Petri,¹² Susan Manzi,¹³ Peter K. Gregersen,³ Timothy W. Behrens,² and Lindsey A. Criswell¹^*

¹Rosalind Russell Medical Research Center for Arthritis, Department of Medicine, University of California San Francisco, San Francisco, California, United States of America

²ITGR Human Genetics, Genentech, South San Francisco, California, United States of America

³Robert S. Boas Center for Genomics and Human Genetics, Feinstein Institute for Medical Research, Manhasset, New York, United States of America

⁴Department of Biostatistical Sciences, Wake Forest University Health Sciences, Wake Forest, North Carolina, United States of America

⁵Department of Medicine, Keck School of Medicine, University of Southern California, Los Angeles, California, United States of America

⁶Department of Human Genetics, University of Pittsburgh Graduate School of Public Health, Pittsburgh, Pennsylvania, United States of America

⁷Arthritis and Immunology Research Program, Oklahoma Medical Research Foundation, Oklahoma City, Oklahoma, United States of America

⁸Center for Genomics and Oncological Research, Pfizer-University of Granada-Junta de Andalucia, Granada, Spain

⁹Department of Medicine, University of California Los Angeles, Los Angeles, California, United States of America

¹⁰U.S. Department of Veterans Affairs Medical Center Cincinnati, Ohio, United States of America

¹¹Cincinnati Children's Hospital Medical Center, Cincinnati, Ohio, United States of America

¹²Division of Rheumatology, Johns Hopkins University School of Medicine, Baltimore, Maryland, United States of America

¹³Allegheny-Singer Research Institute, Pittsburgh, Pennsylvania, United States of America

Mark I. McCarthy, Editor

University of Oxford, United Kingdom

* E-mail: lindsey.criswell/at/ucsf.edu

Conceived and designed the experiments: KET SAC RRG PKG TWB LAC. Analyzed the data: KET CDL. Contributed reagents/materials/analysis tools: RRG WAO COJ PKG TWB. Wrote the paper: KET. Performed genotyping: ATL. Recruited subjects and collected data: MIK MEAR BPT KLM PMG JBH SM MP.

Received July 12, 2010; Accepted January 13, 2011.

This article has been cited by other articles in PMC.

Abstract

Systemic lupus erythematosus (SLE) is a genetically complex disease with heterogeneous clinical manifestations. Recent studies have greatly expanded the number of established SLE risk alleles, but the distribution of multiple risk alleles in cases versus controls and their relationship to subphenotypes have not been studied. We studied 22 SLE susceptibility polymorphisms with previous genome-wide evidence of association (p<5×10⁻⁸) in 1919 SLE cases from 9 independent Caucasian SLE case series and 4813 independent controls. The mean number of risk alleles in cases was 15.1 (SD 3.1) while the mean in controls was 13.1 (SD 2.8), with trend p

4×10⁻¹²⁸. We defined a genetic risk score (GRS) for SLE as the number of risk alleles with each weighted by the SLE risk odds ratio (OR). The OR for high-low GRS tertiles, adjusted for intra-European ancestry, sex, and parent study, was 4.4 (95% CI 3.8–5.1). We studied associations of individual SNPs and the GRS with clinical manifestations for the cases: age at diagnosis, the 11 American College of Rheumatology classification criteria, and double-stranded DNA antibody (anti-dsDNA) production. Six subphenotypes were significantly associated with the GRS, most notably anti-dsDNA (OR_high-low

2.36, p

9e−9), the immunologic criterion (OR_high-low

2.23, p

3e−7), and age at diagnosis (OR_high-low

1.45, p

0.0060). Finally, we developed a subphenotype-specific GRS (sub-GRS) for each phenotype with more power to detect cumulative genetic associations. The sub-GRS was more strongly associated than any single SNP effect for 5 subphenotypes (the above plus hematologic disorder and oral ulcers), while single loci are more significantly associated with renal disease (HLA-DRB1, OR

1.37, 95% CI 1.14–1.64) and arthritis (ITGAM, OR

0.72, 95% CI 0.59–0.88). We did not observe significant associations for other subphenotypes, for individual loci or the sub-GRS. Thus our analysis categorizes SLE subphenotypes into three groups: those having cumulative, single, and no known genetic association with respect to the currently established SLE risk loci.

Author Summary

Systemic lupus erythematosus is a chronic disabling autoimmune disease, most commonly striking women in their thirties or forties. It can cause a wide variety of clinical manifestations, including kidney disease, arthritis, and skin disorders. Prognosis varies greatly depending on these clinical features, with kidney disease and related characteristics leading to greater morbidity and mortality. It is also complex genetically; while lupus runs in families, genes increase one's risk for lupus but do not fully determine the outcome. The interactions of multiple genes and/or interactions between genes and environmental factors may cause lupus, but the causes and disease pathways of this very heterogeneous disease are not well understood. By examining relationships between the presence of multiple lupus risk genes, lupus susceptibility, and clinical manifestations, we hope to better understand how lupus is triggered and by what biological pathways it progresses. We show in this work that certain clinical manifestations of lupus are highly associated with cumulative genetic variations, i.e. multiple risk alleles, while others are associated with a single variation or none at all.

Introduction

Systemic lupus erythematosus (SLE) is a debilitating autoimmune disease affecting multiple organ systems and characterized by the production of multiple autoantibodies. It is genetically complex, with an estimated sibling risk ratio An external file that holds a picture, illustration, etc.
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_s of 8–29 and heritability greater than 66% [1]. It is also an extremely heterogeneous disease, with patients meeting any 4 out of 11 American College of Rheumatology (ACR) criteria – which include such disparate manifestations as renal disease, arthritis, hematologic disorders, and skin manifestations – classified as having SLE. As these disparate manifestations have great impact on the disease course, understanding their specific genetic etiology is of prime importance.

Until 2008, only a handful of genetic loci affecting SLE susceptibility had been identified and reproduced (e.g. HLA-DRB1, FcγR2A, PTPN22, IRF5, STAT4) via candidate-gene studies [2]. The advent of genome-wide single nucleotide polymorphism (SNP) genotyping technology and subsequent recent genome-wide association studies (GWAS) have greatly expanded the number of established SLE risk alleles [3]–[7] to over twenty; most are located in immune-related pathways such as antigen presentation, B- and T-cell receptor signaling, and interferon signaling [2]. Furthermore, few relationships between SLE clinical manifestations and individual risk alleles have been reported, such as the STAT4 gene variant rs7574865 with production of antibodies to double-stranded DNA (anti-dsDNA) and age at diagnosis [8] and the association between anti-dsDNA production and the HLA-DRB1 *1501 (DR2) allele [9].

However, the distribution of multiple risk alleles in affected and unaffected individuals and the relationship of this distribution to clinical manifestations of SLE have not been studied. Recent studies have begun to characterize cumulative associations of multiple risk alleles for other diseases using a variety of techniques [10]–[12]. In this work, twenty-two genetic variants with p<5×10⁻⁸ in recent studies [6] were chosen for a composite genetic risk score (GRS) for SLE. We investigated the risk alleles and GRS with two goals: first to further characterize SLE susceptibility, and then to investigate relationships with SLE subphenotypes – namely the 11 American College of Rheumatology (ACR) classification criteria [13], [14], anti-dsDNA production (a subset of the ACR immunologic criterion), and age at diagnosis. Our access to a large SLE case collection with clinical and genetic data provided an opportunity to analyze the risk alleles and subphenotypes comprehensively, both as individual alleles and with the genetic risk scores.

Results

Case and control cohorts

A total of 1919 SLE cases and 4813 healthy controls obtained from two SLE GWAS (referred to as “parent studies”) were analyzed. Subjects by parent study and source are shown in Table S1. Parent study 1 contained 1295 cases and 3334 controls genotyped on the Illumina 550K panel as part of an SLE GWAS [3] including 1722 controls from iControlDB (Illumina, www.illumina.com). Parent study 2 contained 624 cases and an additional 337 non-overlapping controls genotyped on the Illumina 317K panel as part of a SLEGEN Consortium GWAS [4], with missing 550K SNPs imputed (see Methods). As many controls were removed from parent study 2 due to overlap with parent study 1, we added an additional 1142 healthy controls from the breast cancer study conducted by CGEMS (Cancer Genetics Markers of Susceptibility, cgems.cancer.gov) [15] genotyped on the Illumina 550K to the parent study 2 dataset.

Individual SLE risk alleles

Twenty-two established SLE risk SNPs with reported genome-wide levels of significance (p<5×10⁻⁸) in at least one study are shown in Table 1, along with their adjusted OR for the cohorts studied here. Associations ranged from OR

1.94 (95% CI 1.75–2.16, p

9×10⁻³⁴) for the HLA-DR3 tag SNP to OR

0.92 (95% CI 0.84–1.00, p

0.055) for BANK1. We also examined all possible 2×2 interactions of the 22 risk alleles in both SLE cases versus controls and in case-only analyses for each subphenotype. No combination was significant using a false discovery rate (FDR) [16] threshold of 5% to account for multiple testing of 22² combinations. For the case-control analysis, the most significant interactions were rs1801274-rs6445975 (FCγR2A -PXK, unadjusted p

0.0022) and rs2187668-rs10488631 (HLA-DR3-IRF5, adjusted p

0.0043). In subphenotype associations, the most significant interactions were rs2431099-rs2187668 (PTTG1-HLA-DR3, unadjusted p

0.0015) for photosensitivity and rs2327832-rs2248932 (TNFAIP3-BLK, unadjusted p

0.0029) for anti-dsDNA production.

Table 1

Twenty-two SNPs used to compute the genetic risk score (GRS), with adjusted odds ratios for the current study.

Genetic risk score

The twenty-two established risk variants were used for two genetic risk measures: a simple count of the number of risk alleles, and a composite genetic risk score (GRS), defined as the summation of SLE risk alleles with each weighted by its SLE odds ratio (OR). Figure 1A shows the distribution of the number of risk alleles in cases versus controls; the mean in cases was 15.1 (SD 3.1) while the mean in controls was 13.1 (SD 2.8), with a trend p

4×10⁻¹²⁸. Figure 1B shows the density of the continuous GRS score, with mean of 18.8 (SD 4.0) in cases and 16.2 (SD 3.6) in controls. The SLE OR, adjusted for intra-European ancestry, sex, and parent study, for high-low GRS tertiles was 4.4 (95% CI 3.8–5.1). Figure 2 shows adjusted ORs for intervals of the GRS compared to a GRS range of 15–17.5. For example, the OR for SLE having GRS>25 versus the reference group was 8.9 (95% CI 5.9–13.2), while the OR for SLE having GRS<10 was 0.29 (95% CI 0.17–0.48).

Figure 1

Distributions of the number of risk alleles and genetic risk score (GRS) by disease status, anti-dsDNA status, and age at diagnosis high-low tertiles.

Figure 2

Odds ratio of GRS intervals, adjusted for four principal components, two parent studies, and gender.

We used receiver operating characteristic (ROC) curves to compare the GRS versus the number of risk alleles as predictors of SLE. The GRS was a significantly better predictor: the area under the curve (AUC) for the GRS was 68.9% (95% CI 67.5%–70.3%) versus 67.9% (95% CI 66.4%–69.3%) for the number of risk alleles, p

3×10⁻¹⁴.

Associations with subphenotypes

As expected, many SLE subphenotypes were significantly correlated, shown in Table S2. The strongest is between anti-dsDNA production and the immunologic disorder classification criterion (r

0.62), as anti-dsDNA is one of many antibodies that fulfill the immunologic criterion. Correlation between anti-dsDNA production and renal disease (r

0.26) and between malar rash and photosensitivity (r

0.18) is also observed. Many of the classification criteria have an inverse association with age at diagnosis, with renal disease showing the strongest inverse correlation (r

−0.27).

Individually, several risk alleles were associated with SLE subphenotypes. Table 2 shows those that were significant at an FDR level of 5%. As reported previously [8], the STAT4 SNP rs7574865 was associated with anti-dsDNA antibody production and early age at diagnosis. We also observed associations between the HLA-DR3 tagging allele and anti-dsDNA production (OR

1.37, 95% CI 1.14–1.65), as well as renal disorder (OR

1.37, 95% CI 1.14–1.64). In addition, we observed significant associations for anti-dsDNA production with ITGAM (OR

1.32, 95% CI 1.09–1.59) and UBE2L3 (OR

1.31, 95% CI 1.09–1.56), arthritis with ITGAM (OR

0.72, 95% CI 0.59–0.88), and immunologic disorder with KIAA1542 (OR

0.79, 95% CI 0.68–0.92) and UHRF1BP1 (OR

1.25, 1.08

1.44), with OR<1 indicating protective effects of the minor allele.

Table 2

Top associations between single risk alleles and subphenotypes.

Next, we considered associations between subphenotypes and cumulative risk alleles. Six subphenotypes were associated with the number of risk alleles (not shown) and the GRS at p<0.05, as shown in Table 3 for the GRS as a continuous score and as a comparison between the highest and lowest tertiles. For all of these subphenotypes, the GRS was slightly more strongly associated than the number of risk alleles (not shown). The strongest associations were between the continuous GRS and anti-dsDNA production (p

9×10⁻¹²), immunologic disorder (p

4×10⁻⁹), and age at diagnosis (continuous, p

9×10⁻⁷). Corresponding ORs for high-low GRS tertiles, respectively, were OR_anti-dsDNA

2.36 (95% CI 1.76–3.16, p

9e−9), OR_immunologic

2.23 (1.64–3.03, p

3e−7), and OR_age

−2.68 (−4.42–0.94, p

0.0026). For the dichotomized age at diagnosis<34 years, OR_age34

1.45 (1.11–1.90, p

0.0060). Figure 1C and 1E shows the distribution of the number of risk alleles in anti-dsDNA positive versus negative SLE subjects and high versus low tertiles of age at diagnosis, respectively. Figure 1D and 1F shows the corresponding density curves for the GRS. We tested these associations both with observations dropped for missing disease duration and on the full dataset using multiple imputation (see Methods) with very similar results (Table S3).

Table 3

Association of GRS with SLE subphenotypes.

However, since the SLE risk alleles include SNPs which may not be associated with subphenotypes, or may have different effect sizes than for SLE susceptibility, the number of risk alleles and the GRS may be under-powered to detect cumulative associations with subphenotypes. To more fully address the question of which subphenotypes have evidence of association with cumulative SLE risk alleles, we used a discovery-replication approach to develop a subphenotype-specific genetic risk score, sub-GRS, for each subphenotype (see Methods) containing a subset of the SLE risk alleles, weighted by the subphenotype odds ratio in parent study 1. The number of SNPs was chosen to optimize the association with the subphenotype, in contrast to the GRS which contains all SLE risk SNPs regardless of the subphenotype associations. Table 4 shows association results for those sub-GRS that have replication p< 0.1 in parent study 2 and ORs in the same direction. ORs are standardized for comparison since the differing number of SNPs and different weights cause each sub-GRS to be on a different scale. Note that all of these subphenotypes were also significantly associated with the SLE GRS, indicating that we did not miss any cumulative associations when testing the SLE GRS. Note that for renal disease, the sub-GRS contained only a single SNP, showing that the top SNP (tagging the HLA-DR3 allele) was more significant than cumulative effects with additional risk alleles. For arthritis, although ITGAM was associated with arthritis in the joint data (FDR p

0.038 in Table 2) and the discovery set (unadjusted p

0.00056), ITGAM and the sub-GRS had p>0.1 in the replication set.

Table 4

Logistic regression results for sub-GRS and SLE GRS, for subphenotypes with sub-GRS p<0.1 in replication set (study 2).

Finally, we considered the predictive capability of the sub-GRS and GRS for associated subphenotypes. We compared ROC curves for four predictive models for each of the six subphenotypes as shown in Table 5. When adding the sub-GRS (model 4) to a model containing only (commonly-available clinical data) disease duration and sex (model 1), the area under the ROC curve was significantly improved (p<0.05) for all of these subphenotypes except renal disease. It was also significantly better than adding only the top single-locus association (model 2). When comparing model 4 to a model containing sex, disease duration, and the SLE GRS, it was only significantly better (p

0.020) for anti-dsDNA production. Figure 3 shows the ROC curves for these four models for anti-dsDNA production.

Table 5

Area under curve for four models.

Figure 3

ROC curves for four anti–dsDNA models.

Discussion

In a large collection of SLE cases and controls, we investigated the relationship between 22 risk alleles, considered individually and as cumulative genetic risk scores, with SLE susceptibility and specific SLE manifestations. It is important to understand the etiology of SLE subphenotypes, since different subphenotypes of SLE have differential morbidity and mortality, and appear likely to have different underlying etiologies as well. We believe that a more clear understanding of which, if any, genes affect each subphenotype may help lead to a better understanding of SLE disease mechanisms.

We defined a genetic risk score, the GRS, as a summation of SLE risk alleles with each allele unit multiplied by the SLE OR for that allele. This is similar to the weighted “wGRS” defined by Karlson et al [11] for rheumatoid arthritis, except that we use the OR directly rather than its logarithm to be on a scale more similar to the number of risk alleles; the use of 22 risk alleles in both is coincidental. While the number of risk alleles is more intuitive and easier to visualize, the GRS has a wider range and variance and a stronger correlation with SLE susceptibility and subphenotypes. When applied to subphenotypes, the GRS may lose power due to unassociated or improperly weighted SNPs. For this reason we also modeled subphenotype-specific genetic risk scores (sub-GRS) with subsets of SNPs determined using a discovery-replication approach. While the association of these scores in our overall dataset was likely to be inflated since a substantial subset of the data was used to determine the ranking and weighting of the composite SNPs, the odds ratios in our replication set were similar or slightly higher than for the SLE GRS.

It should be noted that many of the SLE risk alleles were discovered using subjects in our study; thus our odds ratios may be an overestimate of the actual odds ratios (“winner's curse”) resulting in over-weighting in the GRS for some SNPs. On the other hand, it is likely that many of these SNPs are not the causal variants but markers in LD. In that case, their effect sizes for SLE susceptibility and/or subphenotype associations would be underestimated, causing the GRS and/or sub-GRS scores to be underweighted and under-associated. Also, in some cases we were not able to use directly-genotyped SNPs at exactly the risk locus previously identified in the literature. Three SNPs were imputed in the SLEGEN dataset (Illumina 317K versus 550K, see Table S1), and for 6 SNPs we used a proxy. Use of proxy and/or imputed SNPs may have given us lower power to detect associations if those SNPs were not as accurate or highly associated; however we believe accuracy was assured by high thresholds for imputation inclusion (see Methods) and proxy SNP selection (r²≥0.8). Also, while multiple signals have been implicated in the TNFAIP3 region [5], [17], we were only able to include one locus with a suitable match in our data. Another potential limitation of the GRS is lack of modeling interactions between SNPs. We tested for all 2×2 interactions between the 22 SNPs in our data with no results being significant after multiple-testing correction; however we may have lacked the statistical power to detect such interactions given our sample size.

Our analyses used HLA-DRB1 tagging SNPs for the DRB1*0301 (DR3) and DRB1*1501 (DR2) alleles rather than direct HLA-DRB1 genotyping data. Our resulting ORs were lower than those in the literature and therefore may underestimate the GRS. We performed sensitivity analyses with a subset of our cases having 4-digit HLA-DRB1 typing (n

716) and a subset of controls having mixed 2- and 4-digit typing (n

1414). Removing ambiguous 2-digit types, there was 98.9% agreement of the DR3 classification (as 0/1/2 alleles) and 98.2% agreement for DR2. We were not able to assess case-control ORs using this data due to the differential typing; however we tested our DR3 associations with anti-dsDNA production and renal subphenotypes, and observed nearly identical ORs and significance compared to the tag SNPs using the same subset of subjects (data not shown).

We have shown that a subset of SLE clinical manifestations – immunological disorder including anti-dsDNA production, renal disease, age at diagnosis, hematologic disorder, and oral ulcers – are strongly associated with the number of risk alleles and the GRS. For most of these, the GRS was much more highly associated than any single locus, with the exception of renal disease and the HLA-DRB1 *0301 (DR3) allele, which is stronger than the GRS signal (and equivalent to the sub-GRS as it had only a single allele). For arthritis, there was no association with the GRS, but there is evidence for a protective effect of the ITGAM locus. For other manifestations, such as malar rash and serositis, there were no significant associations with either the GRS, sub-GRS, or with single loci. This led to our categorization of SLE manifestations into those that are: a) influenced by cumulative effects of multiple known genes, b) influenced primarily by a single gene out of the currently-established risk loci, and c) thus far not appearing to be strongly influenced by genetics (Figure 4). Anti-nuclear antibody production was not included in this characterization as it was present in almost all SLE patients (95.9% of our subjects, Table S4); it is also possible that some associations were not evident due to lack of power for less-frequent manifestations, such as discoid rash and neurologic disorder.

Figure 4

Categorization of SLE subphenotypes by strongest association with currently known susceptibility loci: genetic risk score, single locus, or none.

Strengths of this study include the large sample size and availability of clinical data for the SLE cases. Although there are potential issues of differing clinical evaluation at different sites and comprehensive follow-up after DNA collection, we expect the standardized ACR criteria to be highly consistent; furthermore we expect that any misclassification would be random with respect to genotype and therefore bias our results towards the null. One related issue was the large number of cases lacking data for disease duration. In general, we took a conservative approach and did not include observations that did not have disease duration information when disease duration was found to be associated with subphenotypes; for a subset of analyses, we also utilized single and/or multiple imputation on the entire dataset and observed similar results.

A limitation of this and most other recent studies of SLE genetics is that it contains only subjects of European ancestry, and primarily northern European. The GRS was strongly associated with the first principal component of whole-genome SNPs, which reflects ancestry along the northwest-to-southeast European cline. This is likely to be at least somewhat if not largely due to the fact that these risk alleles have been discovered using mostly subjects of northern European ancestry, and additional risk alleles for other populations have yet to be discovered.

While the GRS was very highly associated with SLE susceptibility, the predictive capability was somewhat modest (AUC for ROC curve 68.9%). For subphenotypes associated with the GRS and sub-GRS, these scores significantly improve prediction over disease duration and gender, but the AUC for these subphenotypes is even more modest (56.0%–65.7%). For renal disease, the GRS did not improve prediction over clinical variables. It will be very interesting to see how such measures will be improved as we obtain additional information on SLE risk. In particular we anticipate that new susceptibility loci will be found as non-northern-Europeans are studied in greater detail. We also anticipate that the locations of current risk loci will be determined more precisely with regional fine mapping, re-sequencing, and functional studies.

Methods

Prior to merging, individual datasets were filtered for individuals with<90% genotyping and SNPs with<90% genotyping, minor allele frequency (MAF)<1%, or HWE p-value<0.00001. SNPs in the 550K but not the 317K platform were imputed in the parent study 2 (SLEGEN) dataset using IMPUTE [18], retaining SNPs with >90% confidence, >90% concordance in two test datasets (500 cases and 500 controls from parent study 1 with known genotypes removed), and >90% imputed genotype rate. In the final merged dataset of genotyped and imputed SNPs, SNPs were again filtered for >90% genotyping (using typed or imputed values). From this dataset, SLE risk SNPs or their proxies were obtained. Out of 22 loci selected for inclusion based on p<5×10⁻⁸ in a previous study [6], 16 were directly genotyped in all of our subjects. Three SNPs were imputed in the SLEGEN dataset, and a proxy SNP (r²>0.8) was found for 6 SNPs using the HapMap (http://www.hapmap.org) CEU population (with one overlap, a proxy SNP imputed in the SLEGEN dataset). Imputed and proxy SNPs are shown in Table S5.

Principal components analysis using EIGENSTRAT [19] was performed using the above merged dataset of directly genotyped SNPs, with SNPs having at least 90% genotyping (thus on both the 317K and 550K platforms). SNPs in regions of known high LD (chr 5: 44–51.5 Mb, chr 6: 25–33.5 Mb, chr 8: 8–12 Mb, chr 11: 45–57 Mb, and chr 17: 40–43 Mb) were removed prior to analysis. Individuals with values more than 6 standard deviations away from the mean of any of the first 10 PCs (n

21) were considered genetic outliers and were removed. Four PCs were used for ancestry adjustment, based on leveling off of the PCA scree plot and due to significant differences between cases and controls for the first 4 PCs.

The GRS was defined as the number of risk alleles at each locus multiplied by the OR for SLE susceptibility in our dataset. For example, two STAT4 risk alleles contribute 2*1.5

3 to the GRS. For a protective SNP, the risk alleles are the major alleles. Since not counting sporadic missing data would underestimate the number of risk alleles, the GRS, and the sub-GRS, we used best-guess imputed missing genotypes (using IMPUTE version 2) for these calculations. The GRS was analyzed both continuously and by comparing the highest and lowest tertiles to aid in interpretation, with comparison of tertiles being a compromise between more extreme tails of the distribution (having less power) and dichotomizing (having less differentiation).

Subphenotypes and covariates studied are shown in Table S4. In each study, subphenotype status was confirmed by chart review. Autoantibody status was determined by chart review and/or serologic testing; subjects were considered auto-antibody positive if there was any positive test indicated in the reviewed medical records or serologic tests. Negative status required that at least one negative test be documented and no positive tests. Positive anti-dsDNA status is a subset of the immunologic criteria; other qualifiers are anti-Sm antibodies or the presence of anti-phospholipid antibodies. Where appropriate, e.g. logistic regression and bar graphs, the age at diagnosis was dichotomized into high-low halves or split into tertiles. For regression, in addition to the ancestry principal components described above, additional covariates were sex, disease duration, and study (two parent studies or eight sources, see Table S1). All subphenotypes were heterogeneous by study source (data not shown).

We first looked at the adjusted association between each outcome and the continuous GRS (Table S3). As we have a high percentage of missing data for disease duration (18.5%, see Table S4), adjustment was done two ways: a) using only the subset of subjects having disease duration, and b) using multiple imputation of the missing disease duration values. Multiple imputation was performed using Stata ICE [20] with predictive matching. Covariates age at diagnosis, study source, and sex were used in the imputation. Differences in results using these methods were very slight for subphenotypes associated with the GRS. We used actual data without imputation in subsequent GRS analyses. For the sub-GRS computations (below), we used single imputation based on the same variables as above.

In subphenotype associations, the SLE GRS may have less power than a risk score which utilizes the SNPs and effect sizes appropriate for that subphenotype. Thus we also tested a subphenotype-specific sub-GRS for each subphenotype, defined via a discovery-replication approach. First, for each subphenotype we used the associations in parent study 1 (the “discovery” study for this analysis) to determine the rank and OR of each risk SNP association with the subphenotype. Then a series of 22 candidate sub-GRS(n) scores were computed incrementally adding in the OR weights by rank, where n is the number of SNPs included. (The first candidate sub-GRS(1) is equal to the top SNP weights, the second candidate sub-GRS(2) adds in the second SNP weights, and so on). The associations in the discovery set for the resulting sub-GRS(n) candidates are shown in Figure 5; p-values are for the likelihood ratio test of differences between models with the sub-GRS(n) plus covariates versus a model with only covariates. This method can accumulate random associations as well, as illustrated for comparison purposes by sample “null” subphenotypes with 50–50 random associations (highest and lowest associations out of ten samples are shown); hence the importance of a discovery-replication approach. Finally the peak association sub-GRS(n) candidate for the replication and discovery sets with the minimum number of SNPs was used as the final sub-GRS for each subphenotype; this assumes that post-peak SNPs in either set are likely to be false positive associations.

Figure 5

Association of subphenotypes with sub-GRS candidates in study 1, by number of included SNPs.

Stata 9.2 [21] was used for regressions and ROC curve analyses. Plink [22] was used for quality control filters, regressions and tests for 2×2 interactions. HelixTree SVS Version 7.2.3 (www.goldenhelix.com) was used for likelihood-ratio tests of logistic regressions of the sub-GRS(n) series. The R programming environment [23] Version 2.11.1 was used for GRS density curves.

Supporting Information

Table S1

Subjects by parent study and source.

(0.05 MB DOC)

Click here for additional data file.^{(46K, doc)}

Table S2

Correlation (r) between SLE subphenotypes.

(0.04 MB DOC)

Click here for additional data file.^{(41K, doc)}

Table S3

Association of GRS with SLE subphenotypes, by missing data method.

(0.04 MB DOC)

Click here for additional data file.^{(40K, doc)}

Table S4

Clinical data summary.

(0.05 MB DOC)

Click here for additional data file.^{(44K, doc)}

Table S5

SNPs of genetic risk score: references, proxies, imputed SNPs, and risk/protective status.

(0.12 MB DOC)

Click here for additional data file.^{(120K, doc)}

Acknowledgments

We would like to thank the many lupus participants and healthy controls for their contributions to our research.

Footnotes

RRG and TWB are full-time employees of Genentech.

Support for this study was provided by the NIH grants RR024130 (SAC), AR043815 (COJ), HL054900 (MIK), HL074165 (MIK), RR020143 (MEA-R), AR043814 (BPT), AR043247 (KLM), AI063274 (PMG), AR052125 (PMG), AR062277 (JBH), RR020143 (JBH), AR042460 (JBH), AI024717 (JBH), AR043727 (MP), AR002213 (SM), AR046588 (SM), AR012256 (PKG), AI095386 (PKG), AI068759 (PKG), AR044804 (LAC), AR02175 (LAC), and AR052300 (LAC). Additional support was provided by the American College of Rheumatology Physician Scientist Development Award (SAC); Wake Forest University Health Sciences Center for Public Health Genomics (CDL), Alliance for Lupus Research (SLEGEN, CDL, COJ, MEA-R, BPT, KLM, PMG, JBH, LAC); the Swedish Research Council, the Swedish Association Against Rheumatism, the King Gustaf Vth-80th-year Jubilee, the grant PS09/00129 Instituto de Salud Carlos III, and the grant PI-0012 from the Consejería de Salud of Andaluca (MEA-R); Lupus Foundation of Minnesota (KLM); Mary Kirkland Scholar Awards (JBH, LAC); US Department of Veterans Affairs (JBH); grant UL1 RR 025005 from the National Center for Research Resources (MP); and the Rosalind Russell Medical Research Center for Arthritis (LAC). These studies were performed in part in the University of California San Francisco, General Clinical Research Center (Moffitt Hospital), the Johns Hopkins University General Clinical Research Center, and the University of Pittsburgh General Clinical Research Center with funds provided by the U.S. Public Health Service National Center for Research Resources, M01 RR-00079, M01 RR-00052, and M01-RR000056. The Broad Institute Center for Genotyping and Analysis is supported by grant U54 RR020278-01 from the National Center for Research Resources. The funders had no role in the study design, data collection and analysis, decision to publish, or preparation of the manuscript.

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