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1 Cancer Epidemiology Centre, The Cancer Council Victoria; 2 Centre for Molecular, Environmental, Genetic, and Analytic Epidemiology and 3 Department of Pathology, University of Melbourne, Melbourne, Victoria, Australia; 4 Cancer Research Program, Garvan Institute of Medical Research, St Vincent's Hospital, Sydney, New South Wales, Australia; 5 Dame Roma Mitchell Cancer Research Laboratories, University of Adelaide; 6 Hanson Institute, Adelaide, South Australia, Australia; and 7 IARC, Lyons, France
Requests for reprints: Graham G. Giles, Cancer Epidemiology Center, Cancer Control Research Institute, Anti-Cancer Council of Victoria, 100 Drummond Street, Carlton, Victoria 3053, Australia. Phone: 61-3-9635-5155; Fax: 61-3-9635-5330. E-mail: graham.giles{at}cancervic.org.au
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Abstract |
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0.2) and rs925013 (1.1, 1.2, and 1.5 ng/mL, respectively; all P > 0.1). For rs925013, our study provides good evidence of association with prostate cancer risk, marginal evidence of association with survival, and little evidence of detectable association with circulating PSA levels in controls. We found no evidence of an independent association between rs266882 and any of the outcomes. The genotypes and haplotypes studied might be associated with the PSA gene function or be in linkage disequilibrium with other unmeasured and functional variants in the PSA or other genes. (Cancer Epidemiol Biomarkers Prev 2006;15(6):1142-7) |
Introduction |
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The PSA gene contains several androgen-responsive elements, which mediate the transcriptional response of the activated androgen receptor that plays a key role in prostate cancer (1, 2). One androgen-responsive element is located in the proximal promoter at –156 to –170 bp from the transcriptional start site of the gene and contains a polymorphic locus (rs266882) at –158 (A to G substitution). Some studies (3-8), but not others (9-12), found that this polymorphism is associated with the development of prostate cancer or circulating PSA levels. A recent study further characterized the PSA gene for polymorphisms and identified several sequence variants, including one further upstream of the PSA promoter at position –4643 (rs925013; ref. 13). This A to G substitution was associated with increased serum PSA levels and transcriptional activity of PSA promoter constructs (13).
We tested the hypothesis that the two single-nucleotide polymorphisms in the PSA gene, rs925013 and rs266882, are associated with circulating levels of PSA, prostate cancer risk, and the risk of dying of the disease using blood samples collected during a large population-based case-control study of prostate cancer.
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Materials and Methods |
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Informed consent was obtained from all study participants. Blood samples were available from 831 cases (79% of participants) and 738 controls (70%). A detailed description of participant characteristics has been published (18). Vital status as of December 31, 2004 and cause of death were determined for the 640 cases in the Melbourne arm of the study by linking these cases to the Victorian Registry of Births, Deaths, and Marriages. During an average follow-up of 8.2 years, 68 cases (11%) were found to have died from prostate cancer.
Genotyping of the PSA Gene Polymorphisms
Genomic DNA was extracted from whole blood and genotyped blind to case-control status. The rs266882 polymorphism was genotyped using PCR-based denaturing gradient gel electrophoresis and the Ingeny phorU-2 denaturing gradient gel electrophoresis system (Ingeny International, Goes, the Netherlands; http://www.ingeny.com). The following primer pair, forward (5'-GTGCATCCAGGGTGATCTAGTA-3') and reverse (5'-CTGCTGGAGGCTGGACAAC-3'), was used to amplify the 141-base fragment. To prevent complete strand dissociation during electrophoresis in a 9% polyacrylamide gel containing a 40% to 80% urea and formamide, denaturing gradient, a 40-base GC-clamp was added to the 5'-end of the forward primer. Conditions of amplification, thermal cycling, and gel conditions are available on request. The rs266882 AG genotype was identified as four denaturing gradient gel electrophoresis bands (two homoduplex and two heteroduplex), whereas the AA and GG genotypes were identified as a single upper and lower homoduplex band, respectively (Fig. 1
). A random selection of 143 samples (9%) were regenotyped with a concordance rate of 100%. The rs925013 was genotyped using matrix-assisted laser desorption/ionization time-of-flight mass spectrometry (Compact Sequenom MassArray system, Sequenom, San Diego, CA). Amplification was done using the following primer pair, forward (5'-ACGTTGGATGATAGAGTCAAGAGGGTACAG-3') and reverse (5'-ACGTTGGATGTTGACCCTCTCTTTTAGGGC-3'), whereas the extension reaction was done using the extended primer (5'-TTTCTGACCTCCACCATGA-3') and 1x ACT termination mix (ddATP, ddCTP, ddTTP, and dGTP). Conditions for amplification and the homogenous mass extend reaction are available on request. Purified extended products were dispensed onto a SpectroCHIP and genoptyped using SpectroTYPER on the MassArray. A random selection of 659 samples (42%) were regenotyped with a concordance of 99.4%.
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Statistical Methods
Estimates of allele frequencies and tests of deviation from Hardy-Weinberg equilibrium were carried out using standard procedures based on asymptotic likelihood theory (19). Linkage disequilibrium between the two variants was assessed by using Lewontin's D' (20), and tests for significance were based on asymptotic likelihood theory. Fisher's exact test was used to test for independence between the single-nucleotide polymorphisms and categorized risk factors [i.e., age (<55, 55-64, and 65-69), country of birth (Australia and others), family history of prostate cancer (affected first-degree relatives and no affected relatives), and tumor stage (stage I-IV) and grade (moderate and high)]. Tests for association between genotypes and the various outcome of interest (i.e., prostate cancer risk, disease-specific survival, and circulating levels of PSA) were done under codominant, dominant, and recessive models. Case-control analyses were conducted using unconditional logistic regression (21), and odds ratio (OR) estimates and their 95% confidence intervals (95% CI) were derived under likelihood theory. Adjustment for country of birth, age, history of smoking, history of prostate cancer in first-degree relatives (family history), body mass index, and alcohol consumption did not materially change the OR estimates from the logistic models. Polytomous logistic regression models were used to estimate ORs by tumor stage (dependent variable with three categories: 0, 1, and 2 for controls, stage I-II tumors, and stage III-IV tumors, respectively) and grade (dependent variable with three categories: 0, 1, and 2 for controls, moderate-grade, and high-grade tumors, respectively).
Survival analysis of prostate cancer cases in the Melbourne arm of the study was used to test the possible effect of genotypes on the risk of dying from prostate cancer. For this purpose Cox regression models were used to estimate the hazard ratios (HR) (22), adjusted for tumor grade and stage, country of birth, age, history of smoking, and family history. Factors were considered confounders and therefore included in the Cox models if they changed the HRs of any of the genetic variants by at least 5%. Further adjustment for body mass index and alcohol consumption did not materially change the HR estimates.
Serum PSA levels were highly skewed, so we used linear regression of the log10-transformed levels of PSA in controls to test the possible association between genotypes and circulating PSA. The linear regression models were adjusted for age and laboratory assay and were first fitted using all the controls and then refitted excluding those with PSA levels of >9 ng/mL as in Cramer et al. (13). Results are presented as adjusted back-transformed means (i.e., geometric means) and their corresponding 95% CI derived from the fitted regression models. These statistical analyses were done using Stata/SE 8.2 (Stata Corporation, College Station, TX).
Estimates of haplotype frequency and tests of association between haplotypes and prostate cancer risk and serum PSA levels were done using the suite of routines HaploStats 1.2.1 (http://mayoresearch.mayo.edu/mayo/research/biostat/schaid.cfm) run from the statistical program R 2.1.1 (http://www.r-project.org). These routines use the expectation maximization algorithm to compute maximum likelihood estimates of haplotype probabilities from genetic markers of unknown linkage phase and posterior probabilities that are then incorporated in regression models for binary or normal outcomes to test for association.
The likelihood ratio test was used to test nested hypotheses and the Wald test to assess statistical significance of individual variables. All tests were two sided. Following convention, nominal statistical significance was based on P < 0.05. No attempt was made to adjust for multiple comparisons.
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Results |
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1% of the samples leaving 821 cases and 734 controls with at least one of the two variants determined (812 cases and 713 controls with both variants determined). Only five cases (<1%) and seven controls (1%) were not of Caucasian origin, with the great majority of subjects (98.5%) being born in Australia, British Isles, or Western Europe. Half of the cases were aged between 55 and 64 years (428, 52%), and 111 cases (14%) were aged <55 years. Two hundred and fifty-two cases (31%) had a stage III or stage IV tumor, and 219 cases (27%) were poorly differentiated or had a Gleason score >8. The distribution of the genotypes was consistent with Hardy-Weinberg equilibrium for both loci in cases and controls and in cases and controls combined (all P > 0.1). The two loci were in strong linkage disequilibrium, and D' was virtually identical in cases and controls (D' = 0.89 for cases and controls combined; P < 0.001). There was no evidence of association between either genotype and age, country of birth, and family history of prostate cancer (all P > 0.06).
Genotypes and Prostate Cancer Risk
For the rs925013 variant, the frequency of the G allele was 18% in controls and 23% in cases, and this allele was associated with increased risk of prostate cancer (P = 0.006 and 0.001 from the codominant and dominant models, respectively). The ORs for men heterozygous and homozygous for the G allele relative to men homozygous for the A allele were 1.4 (95% CI, 1.1-1.7; P = 0.003) and 1.5 (95% CI, 0.9-2.4; P = 0.1), respectively (codominant model). The prevalence of the G allele was similar in stage III to IV and stage I to II tumors (42% and 40%, respectively; P = 0.5), and the ORs did not differ significantly by tumor stage (all P > 0.4). The prevalence of the G allele in moderate-grade tumors was higher than in high-grade tumors (43% and 34%, respectively; P = 0.02), and we found marginal evidence that ORs for moderate-grade tumors, ranging from 1.3 to 1.6, were higher than ORs for high-grade tumors (P = 0.04, 0.02, and 0.2 for the codominant, dominant, and recessive models, respectively).
For the rs266882 variant, the frequency of the G allele was 48% in controls and 50% in cases, and there was little evidence of an association with prostate cancer risk from the codominant, dominant, and recessive models (ORs range, 1-1.2; all P 0.1; Table 1
). The prevalence of the G allele in stage III to IV tumors was higher than in stage I to II tumors (81% and 72%, respectively; P = 0.006). The ORs for stage III to IV tumors were higher than unity (OR, 1.3; 95% CI, 0.9-1.9; P = 0.1 and OR, 1.9; 95% CI, 1.3-3; P = 0.002 from the codominant model; OR, 1.5; 95% CI, 1.1-2.2; P = 0.02 from the dominant model; and OR, 1.6; 95% CI, 1.2-2.2; P = 0.004 from the recessive model) and significantly higher (all P < 0.005) than the ORs for stage I to II tumors that ranged from 0.9 to 1 (all P > 0.3). The prevalence of the G allele was similar in moderate- and high-grade tumors (74% and 75%, respectively; P = 0.7; data not shown), and the ORs did not differ significantly by tumor grade (all P > 0.6).
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0.4 from the additive, dominant, and recessive models, respectively). The G/G haplotype was associated with an increased risk of prostate cancer (P = 0.008 and 0.009 from the additive and dominant models). From the additive model, the ORs for men carrying one or two copies of the G/G haplotype compared with men homozygous for the A/A haplotype were 1.3 (95% CI, 1.1-1.6; P = 0.008) and 1.7 (95% CI, 1.1-2.4; P = 0.008), respectively. The G/A haplotype was too rare to estimate its association with prostate cancer risk.
Risk of Dying From Prostate Cancer
Although the overall tests for association between the rs925013 variant and disease-specific survival were not statistically significant (all P 0.1) from the codominant model, the HRs for cases heterozygous and homozygous for the G allele compared with cases homozygous for the A allele were 1.2 (95% CI, 0.7-2.1; P = 0.4) and 2.3 (95% CI, 1-5.6; P = 0.06), respectively. The rs266882 variant was not associated with disease-specific survival of prostate cancer cases (HRs range, 0.9-1.2; all P 0.5; Table 3
).
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0.2 from the codominant, dominant, and recessive models).
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Discussion |
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The main strength of this study is the large number of cases and controls. With 1,553 participants, this is by far the largest study on the association between genetic polymorphisms in the PSA gene and prostate cancer with almost the same number of participants as all the previous studies combined. This is also the first study that followed up cases to test the possible influence of the variants in the PSA gene on survival and the largest among the studies that measured circulating PSA levels in controls to look for clues of the potential biological effects of the PSA gene variants. In addition, haplotype analysis, done using statistical techniques to derive haplotypes from genotype data, provided additional information to the standard genotype analysis, although the added information was relatively limited because the two variants were in linkage disequilibrium. The inclusion of other variants from the series identified by Cramer et al. would have added little information to the haplotype analysis because these variants were all in strong linkage disequilibrium (13).
One weakness of the study is the limited power of the survival analysis, with 68 deaths due to prostate cancer during the follow-up period. Another weakness was that we could not adjust the survival analysis for PSA levels at diagnosis because we had incomplete information on PSA levels at diagnosis for the cases. This adjustment would have allowed us to test whether the increased HRs for cases homozygous for the G allele in rs925013 were due to increased PSA levels at diagnosis. Although we found little evidence of an association between circulating PSA levels and genotypes in controls, we cannot exclude an association between genotypes and PSA levels at diagnosis for cases. A case-control study originally reported that men with longer CAG repeats in the androgen receptor gene and who were also carriers of the G allele in rs266882 of the PSA gene had lower levels of serum PSA (4). We could not test this hypothesis because we did not measure the number of CAG repeats in the androgen receptor gene, but this finding was not replicated in other studies (6, 9). Although we think it unlikely that genotypes were different in respondents and nonrespondents, limited response rates is another weakness of our study.
Although we cannot rule out the role of chance, there are three plausible explanations for the increased risk of prostate cancer for carriers of the G allele in rs925013 and for the possible increased risk of dying of the disease in cases homozygous for the same allele. The first is that rs925013 plays a role in the development of prostate cancer and in its progression from latent to more aggressive forms, independently of circulating PSA levels. The results of the case-control comparison by tumor stage and grade, however, do not provide evidence of a higher prevalence of the G allele in rs925013 in men with stage III to IV or high-grade tumors and do not explain the results from the survival analysis. The second explanation is that the genetic variant is associated with higher levels of PSA, and, therefore, the association between variant and prostate cancer is only a diagnostic artifact. Our lack of association between rs925013 and circulating PSA levels in controls is evidence against this explanation. However, our results are in contrast with the study by Cramer et al. who showed increased promoter activity and increased PSA levels in carriers of the G allele (13). Third, the associations we found may reflect associations with other genetic variants in linkage disequilibrium with rs925013. Cramer et al., for example, showed that other polymorphisms in strong linkage disequilibrium with rs266882 and rs925013 are associated with variation in circulating levels of PSA (13). The two variants might also be in linkage disequilibrium with variants in other genes involved in the pathway to prostate cancer.
Our results are more consistent with the first explanation. Carriers of the G allele in rs925013 would have an increased risk of prostate cancer not because they have higher levels of PSA but for some other reason. Future studies will have to test this hypothesis through a prospective design or by systematically collecting PSA levels at diagnosis before treatment in a large series of prostate cancer cases. A longer follow-up of these cases in the present study will be necessary to confirm the possible association between rs925013 and survival.
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Acknowledgments |
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Footnotes |
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The costs of publication of this article were defrayed in part by the payment of page charges. This article must therefore be hereby marked advertisement in accordance with 18 U.S.C. Section 1734 solely to indicate this fact.
Note: Current address for P. Neufing: Department of Immunology, Flinders Medical Centre, Adelaide, South Australia, Australia.
Received 12/26/05; revised 3/15/06; accepted 4/ 4/06.
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Commentary
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