RESEARCH ARTICLE

Association between CCND1 and XPC polymorphismsand bladder cancer risk: a meta-analysis based on 15case–control studies

Yifei Wang & Zongping Li & Naibo Liu & Guan Zhang

Received: 30 September 2013 /Accepted: 8 November 2013# International Society of Oncology and BioMarkers (ISOBM) 2013

Abstract Perturbations in cell cycle and DNA repair genesmight affect susceptibility to cancer. The aim of this meta-analysis is to generate large-scale evidence to determine thedegree to which common Cyclin D1 (CCND1 ) G870A(dbSNP: rs603965) and xeroderma pigmentosum group C(XPC) Ala499Val (dbSNP: rs2228000) polymorphisms areassociated with susceptibility to bladder cancer. The electronicdatabases PubMed, Embase, Web of Science, and CNKI weresearched for relevant studies (with an upper date limit of July25, 2013). The principal outcome measure for evaluating thestrength of association was crude odds ratios (ORs) along withtheir corresponding confidence intervals (95 %CIs).We foundand reviewed nine case–control studies on CCND1 G870Awith a total of 6,823 subjects and seven studies on XPCAla499Val with a total of 7,674 subjects. Our meta-analysisprovides evidence that the variant genotype of CCND1G870A showed a significant association in the occurrence ofinvasive bladder tumors in former and current smokers. TheXPC Ala499Val polymorphism correlated with significantdifferences between patients and unaffected subjects, butwhen the groups were stratified by ethnicity, the magnitudeof the overall effect was similar only among Caucasian pop-ulations. Results from our meta-analysis support the view thatthe G870A polymorphism may modulate the risk of bladdercancer in conjunction with tobacco smoking and that theAla499Val polymorphism may contribute to the susceptibilityto bladder cancer in Caucasian populations. Our findings,

however, warrant larger well-designed studies to investigatethe significance of these two polymorphisms as markers ofsusceptibility to bladder cancer.

Keywords CCND1 .XPC . Bladder cancer .

Polymorphism .Meta-analysis

Introduction

Bladder cancer remains one of the most frequently occurringmalignant diseases in men, with a 14-fold variation in inci-dence internationally [1]. Currently, multiple etiologies havebeen indicated in the pathogenesis of bladder cancer.Epidemiological studies have demonstrated that tobaccosmoking is the most prominent and well-established riskfactor for bladder cancer, accounting for 50 % of bladdercancer occurrence in men and 20 % in women [2].Occupational exposures to aromatic amines and amides alsoplay important roles in the etiology of bladder cancer [3].Although the precise mechanism for the association has notbeen fully clarified, a common property of these exposures isthat by-products of cellular metabolism lead to DNA damagein the bladder epithelium, which, if left unrepaired, can causemutations and, ultimately, result in the process of carcinogen-esis [4,5]. Tobacco smoking is also an exceptionally richsource of reactive oxygen species (ROS), which can induceDNA damage and also accumulate in the bladder as by-products of the metabolism of chemical carcinogens presentin tobacco smoke [6,7]. In addition, among those exposed,only a fraction ultimately developed bladder cancer, suggest-ing that certain common genetic variants or polymorphismsmay trigger the disease. As a result, the associations betweengenetic polymorphisms in DNA repair and cell cycle genesand bladder cancer susceptibility have received enormousattention in the last decade.

Y. Wang :N. Liu :G. ZhangDepartment of Urology, China-Japan Friendship Hospital,No. 2 East Yinghua Road, Beijing 100029, China

Z. Li (*)Out-patient Department, Equipment Academy of Air Force, No.11An’ningzhuang Road, Haidian District, Beijing 100085, Chinae-mail: zpli_cjfh@163.com

Tumor Biol.DOI 10.1007/s13277-013-1412-9

Functional single nucleotide polymorphisms (SNPs) occur-ring in DNA repair genes may affect protein function and anindividual's capacity to repair damaged DNA. This deficiencyin repair capacity likely leads to genomic instability, birthdefects, cancer, and reduced life span [8]. Cell cycle check-points represent integral components of DNA repair that re-spond to damage and restore DNA structure, thus providing aprimary defense mechanism against mutagenic exposures [9].DNA repair pathways consist of at least four major types ofcorrections to damaged DNA, such as nucleotide excisionrepair (NER), base excision repair (BER), double-strand breakrepair (DSBR), and mismatch repair (MMR) [10,11]. Up to thepresent, more than 100 gene-coded enzymes have been foundin human cells that are implicated in the four major DNA repairpathways [12,13]. Recently, from the mutational screening ofpatients, common non-truncating polymorphisms in DNA re-pair and cell cycle genes have been identified that may also befunctionally associatedwith bladder cancer risk [14]. The genesstudied in relation to bladder cancer mainly include thexeroderma pigmentosum group D (XPD ) and group C(XPC ), X-ray cross-complementing group 1 (XRCC1 ) andgroup 3 (XRCC3 ), apurinic endonuclease 1 (APE1 ), andCyclin D1 (CCND1). Several meta-analyses with pooled datahave shown that XPD Asp312Asn [15,16], XRCC3Thr241Met [17,18], and XPC Lys939Gln [19,20] polymor-phisms may cause greater susceptibility to bladder cancer,while XPD Lys751Gln [16], APE1 T1349G [21], andXRCC1 (Arg194Trp, Arg280His, and Arg399Gln) [22–24]polymorphisms may have no effect on such susceptibility.

The CCND1 gene is located at chromosome 11p13, and itsamplification is frequently detected in the transitional cellcarcinoma (TCC) of bladder cancer [25]. While CCND1 isreported to be highly polymorphic, the G870A (rs603965)polymorphism, as the most extensively studied SNP, has beenshown to increase levels of circulating CCND1 [26]. Theoverexpression of CCND1 induces excessive cellular prolif-eration, one of the features of cancer, suggesting that theCCND1 gene may contribute to the process of genomicinstability during tumor development [27]. The XPC gene,located at chromosome 19q13.3, encodes a component of theNER pathway and plays an important role in DNA repair [28].Two common polymorphisms in XPC are widely studied: (a)a substitution of alanine for valine in codon 499 (Ala499Val,dbSNP: rs2228000) and (b) an A to C transversion in exon 15resulting in a lysine-to-glutamine transition at position 939(Lys939Gln, dbSNP: rs2228001). In recent years, more andmore epidemiological studies have investigated the role ofselected CCND1 G870A and XPC Ala499Val polymor-phisms in susceptibility to bladder cancer and the potentialmodifier role of these two polymorphisms on the effects ofsmoking [29–33]. These studies, however, have proven con-troversial and inconclusive, and thus, larger sample sizes arerequired to evaluate the associations between these two

polymorphisms and risk of bladder cancer. To address theseissues, we carried out a quantitative meta-analysis of all stud-ies relating CCND1 G870A and XPC Ala499Val polymor-phisms and bladder cancer risk.

Material and methods

Literature search strategy

Relevant studies were identified in the PubMed, Embase,Webof Science, and Chinese National Knowledge Infrastructure(CNKI) (last search updated July 25, 2013) related toCCND1G870A and XPC Ala499Val polymorphisms and bladdercancer risk. Literature searches were performed by combiningtext words and MeSH terms, without any language restriction,using the following search phrases: (“urinary bladder neo-plasms” or “bladder cancer” or “bladder tumors”) and (“poly-morphism, genetic” or “single nucleotide polymorphism” or“SNP”) and (“Cyclin D1” or “CCND1” or “xerodermapigmentosum, complementation group C” or “XPC”). Thereferences cited in retrieved articles were also screened inorder to find any studies that may have been missed in thecomputer-assisted search.

Inclusion and exclusion criteria

Studies qualified for inclusion met the following criteria: (a)the studied population was comprised of unrelated individ-uals, (b) distribution of G870A and Ala499Val genotypes wasdetermined in both bladder cancer patients and in concurrentcontrol group of bladder cancer-free subjects, (c) studies useda 95 %CI for odds ratio (OR) or allowed for the possibility ofcalculating these measures, (d) validated genotyping methodswere used, and (e) the genotype distribution of controls was inHardy–Weinberg equilibrium (HWE).

Meta-analyses, case reports, reviews, and editorial articleswere excluded. In addition, study groups with fewer than 30subjects were excluded from our analysis. Family-based stud-ies were also excluded due to their different experimentaldesign. For studies with the same or overlapping case seriesand by the same investigators, we generally retained the mostrecent ones that had the most subjects for meta-analysis. Forthose studies reporting only the allele frequencies in case andcontrol groups, we contacted the corresponding authors byemail to obtain the additional information and clarifications.All disagreements were resolved by discussions and subse-quent consensus.

Data extraction

For each study and study group, we coded information on thefirst author's last name, year of publication, country of study

Tumor Biol.

population, subjects' ethnicities, number of cases and controls,mean age (or range whenever possible) of cases and controls,SNPs investigated, method of genotyping, the genetic poly-morphism(s) assessed, allele/genotype frequency in case andcontrols if available, and evidence of HWE in controls. Inaddition, we compared key study characteristics such as loca-tion, study time, and authorship to determine the existence ofmultiple publications from the same study. Two investigatorsindependently extracted the data using a piloted data standard-ized form and reached a consensus on all classified items. Incases of conflicting evaluations, minor discrepancies wereresolved by consensus with a third investigator's careful reex-amination of the full texts.

Quality assessment of included studies

Methodological quality of included studies was assessed inde-pendently by two investigators based on the modifiedStrengthening the Reporting of Observational Studies inEpidemiology (STROBE) quality score system [34]. The qual-ity scale consists of 40 items with scores ranging from 0 to 40.The eligible studies were classified into three levels based ontheir scores: low quality (0–19), moderate quality (20–29), andhigh quality (30–40). Disagreements on STROBE scores of theincluded studies were resolved through a comprehensive reas-sessment by the authors of this meta-analysis.

Statistical analysis

Crude ORs corresponding to 95%CIswere applied, accordingto the method of Woolf [35], to assess the strength of theassociation between CCND1 G870A and XPC Ala499Valpolymorphisms and bladder cancer risk under five geneticmodels: the allele model (A vs. a), the dominant model(AA + Aa vs. aa), the recessive model (AA vs. Aa + aa), thehomozygous model (AA vs. aa), and the heterozygous model(AAvs. Aa) (A: mutant allele, a: wild allele; 870A and 499Valwere considered as the mutant alleles). Genotype distributionsin the control subjects were tested for HWE using the chi-square goodness of fit test with a P value of <0.05 beingconsidered significant. Heterogeneity was checked by aCochran's Q-statistic, which is also considered significant atP <0.05 [36]. The I2 test was also used to quantify heteroge-neity (ranging from 0 to 100 %) [37]. When a P valueof <0.05 for Q -test or I 2>50 % indicated the existence ofheterogeneity across the studies, a random effects model(DerSimonian–Laird method) was used; otherwise, a fixedeffects model (Mantel–Haenszel method) was applied in ouranalysis.

For each genetic contrast, a subgroup analysis according toethnicity (i.e., Caucasian or Asian) was conducted for theG870A and Ala499Val polymorphisms, while subgroup anal-yses according to tumor stage (i.e., invasive or non-invasive)

and smoking history (i.e., ever-smoker or non-smoker) wereconducted only for the G870A polymorphism since there wasinsufficient data on the Ala499Val polymorphism since therewas only one study providing data on invasive and non-invasive bladder cancer cases. To evaluate the stability of theresults, a one-way sensitivity analysis was conducted by plot-ting the summary OR with each study omitted in turn. Aninverted funnel plot and Egger's linear regression test wereused to provide a diagnosis of publication bias [38]. Thesignificance of the pooled data was determined using the Ztest and a P value of <0.05 (two-sided) was considered asstatistically significant. All the statistical tests for our meta-analysis were conducted using the software Stata version 12.0(Stata Corp, College Station, TX, USA).

Results

The characteristics of included studies

According to our bibliographic search and selection criteria,228 potentially eligible studies were retrieved and 17 case–control studies were considered [29–33,39–50]. Among them,two publications were excluded due to multiple reporting onthe same population [49,50] and 15 case–control studies wereselected for data extraction and assessment [29–33,39–48], aslisted in Table 1. Flow chart of the study selection process andthe specific reasons for any exclusion from this meta-analysisis shown in Fig. 1. All of the studies were published inEnglish. Among the 15 studies, 8 studies investigatedCCND1 G870A polymorphism [29–31,39,40,43,45,47,48],and 7 focused on XPC Ala499Val polymorphism[32,33,41–44,46]. Five of the 15 case–control studies wereconducted with Asian populations, and the remaining 10studies were conducted with Caucasian populations. All in-cluded studies extracted DNA from peripheral blood for pro-filing gene polymorphisms, and ten articles used the classicpolymerase chain reaction–restriction fragment length poly-morphism (PCR-RFLP) to validate genotypes. TaqMan SNPgenotyping assay was used in four articles and MassARRAYwas used in one article. The genetic distributions of the controlgroups of all studies were coherent with the assumption ofHWE (all P >0.05). Methodological quality of included stud-ies was acceptable, and all quality scores of the includedstudies were higher than 20 (moderate–high quality). Thecharacteristics and methodological quality of the includedstudies are summarized in Table 1.

Association between CCND1 G870A polymorphismand susceptibility to bladder cancer

Table 2 shows the results of the combined data of the ninecase–control studies with 3,153 cases and 3,670 controls. It

Tumor Biol.

depicts a plot of ORs (95 %CI) for the risk of developingbladder cancer associated with CCND1 G870A polymor-phism. Since heterogeneity obviously existed under severalgenetic models, the random effects model was applied.Overall, the meta-analysis results revealed that no significantassociation was found between G870A polymorphism andbladder cancer risk regardless of the genetic contrast(P > 0.05 for all comparisons). However, when anethnicity-stratified analysis was performed, a significantlyhigher prevalence of the 870A allele was observed amongAsian populations under all genetic models (for allele model:OR=1.26, 95 %CI 1.13–1.41, P <0.001; dominant model:OR=1.32, 95 %CI 1.09–1.60, P=0.004; recessive model:OR=1.39, 95 %CI 1.17–1.65, P <0.001; homozygous model:OR=1.59, 95%CI 1.26–2.02, P <0.001; heterozygousmodel:OR=1.32, 95 %CI 1.10–1.58, P=0.003), but no evidence ofany significant association was observed in Caucasian popu-lations (P >0.05 for all comparisons) (Fig. 2a). Further sub-group analyses based on the tumor stage of bladder cancer andsmoking history suggested that the A allele of G870A poly-morphism was associated with increased bladder cancer risk,invasive bladder cancer, and a gene dosage effect in smokingpatients (data not shown), while similar findings were notobserved in non-invasive tumors and non-smoker patients(P >0.05 for all comparisons) (Fig. 2a, b).

Association between XPC Ala499Val polymorphismand susceptibility to bladder cancer

The association between the XPC Ala499Val polymorphismand bladder cancer was investigated in seven studies with aT

able1

Characteristicsandmethodologicalq

ualityof

theincluded

studies

Mutationsite(aliasname)

Study(firstauthor,year)

Country

Ethnicity

Cases

[n,age

(years)]

Controls[n,age

(years)]

Male(%

)[case,control]

Detectio

nmethod

Qualityscores

G870A

(Pro241P

ro)

WangL,2002

Japan

Asian

222,mean65.7(SD11.9)

317,mean52.5(SD11.7)

77.5,69.1

PCR-RFL

P25

CortessisVK,2003

USA

Caucasian

515,range25–64

612,NA

78.6,79.9

PCR-RFL

P23

SanyalS,2004

Germany

Caucasian

327,range33–96

246,NA

NA

PCR-RFL

P27

WuXJ,2006

USA

Caucasian

696,mean63.9(SD11.2)

629,mean62.77(SD10.5)

78.45,72.66

PCR-RFL

P24

Chung

CJ,2008

China

Asian

170,mean62.1(SD1.08)

402,mean61.5(SD0.71)

69.9

PCR-RFL

P25

YeY,2008

USA

Caucasian

629,mean64.1(SD11.2)

629,mean62.8(SD10.5)

78.1,72.58

TaqM

an26

Gangw

arR,2010

India

Asian

212,mean58.5(SD12.4)

250,mean56.8(SD10.8)

88.2,86

PCR-RFL

P28

YuanL,2010

China

Asian

402,mean63.9(SD13.0)

402,mean62.9(SD11.8)

82.1,81.1

PCR-RFL

P22

Lin

HH,2011

China

Asian

101,NA

243,NA

NA

PCR-RFL

P25

Ala499V

al(A

499V

)Broberg

K,2005

Sweden

Caucasian

61,N

A155,NA

NA

MassA

RRAY

28Garcia-ClosasMN,2006

Spain

Caucasian

1,150,mean66

(SD10)

1149,N

ANA

TaqM

an27

SakSC

,2006

UK

Caucasian

547,NA

579,NA

NA

TaqM

an29

WuXJ,2006

USA

Caucasian

696,mean63.9(SD11.2)

629,mean62.77(SD10.5)

78.45,72.66

PCR-RFL

P24

Zhu

YM,2007

USA

Caucasian

578,mean63.7(SD10.8)

578,mean63.1(SD10.5)

77.7,77.7

TaqM

an23

deVerdier

PJ,2010

Sweden

Caucasian

311,range34–96

330,range22–89

NA

PCR-RFL

P26

Liu

Y,2012

China

Asian

600,mean61.8(SD13.5)

609,mean61.4(SD14.1)

79.3,80.0

PCR-RFL

P24

NAnotavailable,PCRpolymerasechainreactio

n,RFLP

restrictionfragmentlengthpolymorphism

Fig. 1 Flow diagram of the selection of studies and specific reasons forexclusion from the present meta-analysis

Tumor Biol.

total of 3,767 cases and 3,907 controls. The findings of thismeta-analysis on the correlation between the SNP and bladdercancer risk are presented in Table 3. The random effects modelwas used to combine data since between-study heterogeneityin seven studies was anticipated. The meta-analysis resultsrevealed that carriers with T(Val) allele of Ala499Val poly-morphism significantly elevated bladder cancer risk under theallele model (OR=1.12, 95 %CI 1.04–1.21, P=0.002), reces-sive model (OR=1.42, 95 %CI 1.19–1.69, P <0.001), homo-zygous model (OR=1.46, 95 %CI 1.22–1.76, P <0.001),and in TT vs. CT contrast (OR=1.37, 95 %CI 1.14–1.66,P=0.001) (Fig. 3). Furthermore, separate meta-analyses strat-ified by ethnicity found that the magnitude of the overall effectwas similar in Caucasian populations (for allele model: OR=1.11, 95 %CI 1.03–1.21, P=0.007; recessive model: OR=1.46, 95%CI 1.19–1.78,P <0.001; homozygousmodel: OR=1.48, 95 %CI 1.21–1.82, P <0.001; heterozygous model:OR=1.43, 95 %CI 1.16–1.76, P=0.001), while the favorabletrend disappeared in Asian populations (P >0.05 for allcomparisons).

Sensitivity analysis and publication bias

Sensitivity analyses were performed by sequentially removingindividual studies and compiling cumulative statistics on al-lele comparison of all subjects and subgroups. As shown inFig. 4, the pooled ORs of G870A and Ala499Val polymor-phisms were not influenced by the results of any individualstudy (data not shown), suggesting that the studies used werestatistically accurate. Furthermore, Begg's funnel plot andEgger's linear regression test were performed on the metadatato assess the publication bias of individual studies. Funnelplots of the allele comparison in the OR analysis for CCND1G870A and XPC Ala499Val (Fig. 5) and Egger's test providedno evidence of publication bias (comparing A vs. G forG870A: t =0.48, P =0.564; comparing T vs. C forAla499Val: t =0.52, P=0.664).

Discussion

Bladder cancer is one of the most common malignancies ofthe urinary tract, with 386,300 new cases and 150,200 deathsfrom bladder cancer in 2008 worldwide [1]. Cigarettesmoking and occupational exposures are the predominant riskfactors for bladder cancer [51]. However, among the exposed,only a fraction actually develops bladder cancer during theirlifetime, revealing variations in both environmental carcino-genic exposures and individual susceptibility to bladder car-cinogenesis [52]. Accumulating evidence indicates that a re-duced DNA repair capacity due to functional SNPs occurringin DNA repair genes is associated with many cancer risks[40,53]. Thus, the associations between the CCND1 G870AT

able2

Meta-analysisof

theassociationbetweenCCND1G870A

polymorphism

andbladdercancer

risk

Subgroups

No.of

study

(case/control)

Aallelevs.G

allele

(allelemodel)

GA+AAvs.G

G(dom

inantm

odel)

AAvs.G

G+GA

(recessive

model)

AAvs.G

G(hom

ozygousmodel)

AAvs.G

A(heterozygousmodel)

OR

95%CI

PPh

OR

95%CI

PPh

OR

95%CI

PPh

OR

95%CI

PPh

OR

95%CI

PPh

Ethnicity

Asian

5(1,107/1,1614)

1.26

1.13–1.41

<0.001

0.580

1.32

1.09–1.60

0.004

0.234

1.39

1.17–1.65

<0.001

0.636

1.59

1.26–2.02

<0.001

0.368

1.32

1.10–1.58

0.003

0.545

Caucasian

4(2,046/2,056)

0.93

0.85–1.02

0.107

0.735

0.88

0.77–1.00

0.052

0.806

0.95

0.82–1.11

0.551

0.869

0.88

0.74–1.05

0.152

0.769

1.00

0.85–1.18

0.971

0.953

Tum

orstage

Invasive

5(478/1,824)

1.21

1.01–1.42

0.034

0.043

1.24

0.97–1.58

0.082

0.065

1.46

1.12–1.91

0.005

0.175

1.28

0.59–2.77

0.524

<0.001

1.48

1.11–1.98

0.007

0.409

Non-invasive

5(938/1,824)

1.05

0.92–1.20

0.436

<0.001

1.09

0.91–1.32

0.334

0.116

1.12

0.91–1.39

0.283

0.001

0.66

0.28–1.57

0.351

<0.001

0.99

0.50–1.98

0.997

<0.001

Smokinghistory

Ever-sm

oker

4(837/748)

1.45

1.15–1.84

0.002

0.245

1.58

1.25–2.01

<0.001

0.917

1.61

1.21–2.14

0.001

0.195

1.96

1.38–2.80

<0.001

0.395

1.44

1.06–1.96

0.018

0.198

Non-smoker

4(442/903)

0.87

0.68–1.11

0.266

0.633

0.95

0.72–1.25

0.698

0.285

0.89

0.59–1.35

0.599

0.679

0.76

0.46–1.25

0.280

0.617

0.98

0.63–1.52

0.931

0.748

Overall

9(3,153/3,670)

1.08

0.96–1.22

0.187

0.005

1.00

0.90–1.12

0.950

0.023

1.15

0.99–1.34

0.068

0.092

1.18

0.93–1.50

0.169

0.005

1.13

1.01–1.28

0.051

0.412

ORodds

ratio

,95%CI95

%confidence

interval,P

hPvalueof

theheterogeneity

test

Tumor Biol.

Overall (I2 = 55.0%, P = 0.023)

Ye Y (2008)

Subtotal (I2 = 28.1%, P = 0.234)

Cortessis VK (2003)

Sanyal S (2004)

Subtotal (I2 = 0.0%, P = 0.806)

Gangwar R (2010)

Chung CJ (2008)

Caucasian

Included Studies

Wang L (2002)

Asian

Wu XJ (2006)

Yuan L (2010)

Lin HH (2011)

1.00 (0.90, 1.12)

0.83 (0.65, 1.06)

1.32 (1.09, 1.60)

0.98 (0.76, 1.27)

0.83 (0.57, 1.22)

0.88 (0.77, 1.00)

1.03 (0.67, 1.60)

1.14 (0.73, 1.77)

OR (95% CI)

1.21 (0.82, 1.78)

0.84 (0.66, 1.07)

1.53 (1.08, 2.16)

2.58 (1.22, 5.48)

100.00

21.12

28.72

17.53

8.97

71.28

6.15

5.73

Weight %

7.36

22.39

7.91

1.56

10.182 5.48

Overall (I2 = 46.1%, P = 0.054)

Lin HH (2011)

Wang L (2002)

Invasive

Subtotal (I2 = 46.0%, P = 0.116)

Cortessis VK (2003)

Gangwar R (2010)

Subtotal (I2 = 54.9%, P = 0.065)

Wang L (2002)

Yuan L (2010)

Gangwar R (2010)

Cortessis VK (2003)

Yuan L (2010)

Non−invasive

Lin HH (2011)

1.15 (0.99, 1.33)

0.54 (0.29, 1.01)

2.04 (0.96, 4.36)

1.10 (0.91, 1.32)

0.89 (0.62, 1.28)

1.41 (0.84, 2.35)

1.24 (0.97, 1.58)

1.04 (0.69, 1.57)

1.39 (0.94, 2.06)

0.93 (0.49, 1.75)

1.05 (0.77, 1.42)

1.83 (1.09, 3.07)

2.15 (0.73, 6.34)

100.00

7.48

3.19

64.17

18.30

7.42

35.83

13.23

12.72

5.75

23.32

6.95

1.63

10.158 6.34

Included Studies OR (95% CI) Weight %

a

b

Overall (I2 = 42.4%, P = 0.096)

Chung CJ (2008)

Subtotal (I2 = 0.0%, P = 0.917)

Chung CJ (2008)

Yuan L (2010)

Cortessis VK (2003)

Cortessis VK (2003)

Ever smoker

Subtotal (I2 = 20.8%, P = 0.285)

Gangwar R (2010)

Gangwar R (2010)

Yuan L (2010)

Never smoker

1.28 (1.06, 1.53)

0.85 (0.47, 1.53)

1.58 (1.25, 2.01)

1.34 (0.63, 2.85)

1.45 (0.85, 2.45)

0.82 (0.49, 1.38)

1.52 (1.10, 2.11)

0.95 (0.72, 1.25)

1.70 (0.88, 3.30)

0.70 (0.38, 1.30)

1.78 (1.10, 2.86)

100.00

11.24

51.51

5.72

11.16

14.83

27.53

48.49

6.25

11.26

12.00

10.303 3.3

Included Studies OR (95% CI) Weight %c

Fig. 2 Meta-analysis for theCCND1 G870A polymorphismand bladder cancer stratifiedaccording to ethnicity (a),smoking history (b), andhistopathologic classification (c)under the dominant model. ORodds ratio, 95 %CI 95 %confidence interval

Tumor Biol.

and XPC Ala499Val polymorphisms and the risk of bladdercancer have been extensively investigated but haveyielded conflicting results in recent years. In addition, anincreasing body of evidence also suggests that these poly-morphisms may alter the functional properties of DNArepair enzymes. Thus, we have undertaken the presentmeta-analysis to make a more precise evaluation of theassociation between bladder cancer risk and variations inCCND1 and XPC .

The present meta-analysis, including 6,832 subjects (3,153 cases and 3,670 controls) from nine case–controlstudies, explores the correlation of common polymor-phism in CCND1 (G870A) with bladder cancer risk.Overall, the data does not show a marked associationbetween genotypes defined by the G870A polymorphismand occurrence of bladder cancer in any genetic model.When we performed an ethnicity-stratified analysis, weobserved that the G870A AA genotype was associatedwith elevated bladder cancer risk only in Asian popula-tions. It is likely that there is an ethnic difference in theG870A genotype; there is a difference between Asiansand Caucasians in the 870A minor allele frequency: 0.57in Asians and 0.44 in Caucasians. Other than familyhistory, the most well-established risk factor for bladdercancer is tobacco consumption, which contributes a morethan 2.5-fold risk in smokers than in non-smokers [54].Therefore, we also conducted a stratified analysis basedon lifetime smoking history. The data shows that the riskcorrelated with the G870A polymorphism was more evi-dent in current and ex-smokers, suggesting that tobaccosmoke may act synergistically with CCND1 genetic alter-ations to modulate susceptibility to bladder cancer. This

finding may also be biologically plausible since Hechtobserved that cigarette smoke generates ROS productionand induces DNA adducts [55]. Furthermore, the stratifiedanalysis according to histopathologic classification ofbladder cancer revealed that 870A is significantly corre-lated with increased risk of invasive tumors, thus indicat-ing that the difference in the levels of the alternateCCND1 transcripts caused by the A/G polymorphismmay influence the biological behavior of bladder cancer.Moreover, these discrepant results could be due to the factthat invasive tumors have more genetic abnormalities aswell as epigenetic defects than superficial tumors.However, it should be noted that these marginally signif-icant results could be merely a result of chance.

Seven studies investigated the association between XPCAla499Val polymorphism and bladder cancer susceptibilitywith a total of 7,674 subjects (3,767 cases and 3,907 controls).The results of this meta-analysis show that the Ala499Valpolymorphism has a statistically significant association withbladder cancer, especially in Caucasian populations, thoughthe functional explanation for such an association remainsundetermined. Previous studies did not predict the effect ofthe Ala499Val substitution on protein function or structure.Therefore, further studies are warranted to validate these ob-servations using a phenotypic-based approach to investigatethe molecular mechanism of these interactions.

This meta-analysis also has some intrinsic limitations.First, due to insufficient data, a stratified analysis of histopath-ologic classification and smoking history was not conductedfor the Ala499Val polymorphism. Second, the sample size forgene–environment interaction assessment in our study wasrelatively modest, so the power of the association analysis

Table 3 Meta-analysis of the as-sociation between XPCAla499Val polymorphism andbladder cancer risk

NA not available, OR odds ratio,95%CI 95% confidence interval,Ph P value of the heterogeneitytest

Genetic models Subgroup No. of study OR 95 %CI P Ph

(case/control)

T allele vs. C allele Overall 7 (3,767/3,907) 1.12 1.04–1.21 0.002 0.002

(allele model) Asian 1 (600/609) 1.16 0.97–1.37 0.093 NA

Caucasian 6 (3,167/3,298) 1.11 1.03–1.21 0.007 0.001

CT + TT vs. CC Overall 7 (3,767/3,907) 1.12 0.95–1.31 0.169 0.010

(dominant model) Asian 1 (600/609) 1.19 0.95–1.50 0.128 NA

Caucasian 6 (3,167/3,298) 1.11 0.92–1.34 0.289 0.006

TT vs. CC + CT Overall 7 (3,767/3,907) 1.42 1.19–1.69 <0.001 0.080

(recessive model) Asian 1 (600/609) 1.28 0.87–1.88 0.210 NA

Caucasian 6 (3,167/3,298) 1.46 1.19–1.78 <0.001 0.051

TT vs. CC Overall 7 (3,767/3,907) 1.46 1.22–1.76 <0.001 0.027

(homozygous model) Asian 1 (600/609) 1.38 0.92–2.07 0.116 NA

Caucasian 6 (3,167/3,298) 1.48 1.21–1.82 <0.001 0.014

TT vs. CT Overall 7 (3,767/3,907) 1.37 1.14–1.66 0.001 0.285

(heterozygous model) Asian 1 (600/609) 1.19 0.79–1.78 0.388 NA

Caucasian 6 (3,167/3,298) 1.43 1.16–1.76 0.001 0.031

Tumor Biol.

Overall (I2 = 70.9%, P = 0.002)

Zhu YM (2007)

Sak SC (2006)

Liu Y (2012)

Asian

Subtotal (I2 = 75.6%, P = 0.001)

Garcia−Closas MN (2006)

Wu XJ (2006)

Broberg K (2005)

de Verdier PJ (2010)

Caucasian

1.12 (1.04, 1.21)

0.96 (0.79, 1.17)

1.23 (1.02, 1.48)

1.16 (0.98, 1.37)

1.11 (1.03, 1.21)

1.06 (0.93, 1.21)

0.96 (0.80, 1.16)

1.20 (0.74, 1.94)

1.80 (1.40, 2.31)

100.00

14.25

13.94

17.70

82.30

29.97

15.48

2.09

6.58

10.433 2.31

Included Studies OR (95% CI) Weight %a

Overall (I2 = 46.8%, P = 0.080)

Caucasian

Liu Y (2012)

de Verdier PJ (2010)

Sak SC (2006)

Zhu YM (2007)

Wu XJ (2006)

Asian

Broberg K (2005)

Subtotal (I2 = 54.6%, P = 0.051)

Garcia−Closas MN (2006)

1.42 (1.19, 1.69)

1.28 (0.87, 1.88)

4.06 (1.97, 8.35)

1.64 (1.07, 2.52)

1.27 (0.73, 2.21)

1.27 (0.74, 2.16)

2.00 (0.67, 6.04)

1.46 (1.20, 1.78)

1.15 (0.83, 1.58)

100.00

22.20

4.15

15.96

10.89

11.52

1.96

77.80

33.32

10.12 8.35

Included Studies OR (95% CI) Weight %b

Overall (I2 = 57.9%, P = 0.027)

Wu XJ (2006)

Sak SC (2006)

Garcia−Closas MN (2006)

Zhu YM (2007)

de Verdier PJ (2010)

Subtotal (I2 = 64.8%, P = 0.014)

Asian

Liu Y (2012)

Broberg K (2005)

Caucasian

1.46 (1.22, 1.76)

1.20 (0.70, 2.07)

1.70 (1.10, 2.65)

1.16 (0.84, 1.62)

1.20 (0.69, 2.10)

4.97 (2.38, 10.37)

1.48 (1.21, 1.82)

1.38 (0.92, 2.07)

1.97 (0.64, 6.09)

100.00

12.32

15.86

33.69

11.67

3.76

79.35

20.65

2.05

10.0964 10.4

Included Studies OR (95% CI) Weight %c

Fig. 3 Meta-analysis for theXPC Ala499Val polymorphismand bladder cancer stratifiedaccording to ethnicity (a for allelemodel, b for recessive model, cfor homozygous model). ORodds ratio, 95 %CI 95 %confidence interval

Tumor Biol.

was inevitably low. Third, we obtained no data fromAfrican populations, which thus ought to be studied andincluded in the future. In addition, consistent with theview of bladder cancer as a multifactorial and multistepdisease, it may also be modulated by several other geneticmarkers besides the CCND1 and XPC genes. Thus, ourmeta-analysis emphasizes that consideration of higher-order interactions among multiple risk elements demandfurther study to reveal the genetic influences on bladdercancer susceptibilities. Finally, our meta-analysis is basedon unadjusted OR estimates because no information onpotential confounders, such as age at diagnosis, gender,and family history of malignant disease, was obtained ineither control or case subjects. Notwithstanding the limi-tations listed above, our meta-analysis has some merit. Tothe best of our knowledge, this is the first meta-analysison the correlation between the CCND1 G870A and XPCAla499Val polymorphisms and bladder cancer.Furthermore, although this meta-analysis does not

accommodate all previously published data, such data islimited compared to the evidence here generated.

In summary, our meta-analysis shows that the CCND1G870A polymorphism is associated with an increased riskof bladder cancer and that this association is stronger forever-smokers and cases of invasive bladder cancer. Inaddition, our results also demonstrated that XPCAla499Val polymorphism may contribute to susceptibilityto bladder cancer in Caucasian populations. We have not,however, studied tobacco smoking, which may influenceXPC gene function, and thus, further studies are warrant-ed to validate these observations. Due to the limitationsmentioned above, our findings need to be validated bylarger well-designed studies with more detailed data onsubjects' environmental exposure.

Conflicts of interest None

0.94 1.08 0.96 1.22 1.26

Wang L (2002)

Cortessis VK (2003)

Sanyal S (2004)

Wu XJ (2006)

Chung CJ (2008)

Ye Y (2008)

Gangwar R (2010)

Yuan L (2010)

Lin HH (2011)

Lower CI Limit Estimate Upper CI Limit

a

1.00 1.12 1.04 1.21 1.25

Broberg K (2005)

Garcia−Closas MN (2006)

Sak SC (2006)

Wu XJ (2006)

Zhu YM (2007)

de Verdier PJ (2010)

Liu Y (2012)

Lower CI Limit Estimate Upper CI Limit

b

Fig. 4 Sensitivity analysis of the summary odds ratio coefficientsof G870A and Ala499Val polymorphisms under the allele model.Results were computed by omitting each study in turn. The twoends of the dotted lines represent the 95 %CI (a G870A, bAla499Val)

LogO

R

SE(LogOR)

0 0.2 0.4

−1

−0.5

0

0.5

1

a(G870A: t = 0.48, P = 0.564)

LogO

R

SE(LogOR)0 0.2 0.4 0.6

−1

0

1

2b

(Ala499Val: t = 0.52, P = 0.664)

Fig. 5 Begg's funnel plot of allele comparison for publication bias(a funnel plot for A vs. G allele comparison in G870A polymor-phism, b funnel plot for T vs. C allele comparison in Ala499Valpolymorphism). Each point represents a separate study by the indi-cated association. LogOR natural logarithm of OR, SE (LogOR )standard error of LogOR. Horizontal line indicates the mean mag-nitude of the effect

Tumor Biol.

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