PPARG Pro12Ala, NFKB1 -94 ins/del, NFKBIA -826C/T, COX-1 (50C>T), and COX-2 (-1195G>A) polymorphisms on prostate cancer risk. The genotypes of the polymorphisms were ascertained in 543 prostate cancer patients and 753 controls through PCR-RFLP and the risk association was evaluated statistically using logistic regression analysis. The NFKB1 -94 polymorphism was shown to decrease prostate cancer risk in both heterozygous and homozygous comparison models (odds ratios of 0.74 (95% CI = 0.58–0.96) (P = 0.02) and 0.57 (95% CI = 0.42–0.78) (P < 0.01), resp.). An opposite finding was observed for COX-2 (-1195) polymorphism (odds ratios of 1.58 (95% CI = 1.15–2.18) (P < 0.01) for heterozygous comparison model and 2.08 (95% CI = 1.48–2.92) (P < 0.01) for homozygous comparison model). No association was observed for other polymorphisms. In conclusion, NFKB1 -94 ins/del and COX-2 (-1195G>A) polymorphisms may be, respectively, associated with decreased and increased prostate cancer risk in the Chinese population.Prostate cancer is a common cancer in men. Genetic variations in inflammatory response genes can potentially influence the risk of prostate cancer. We aimed to examine the association between  Prostate cancer is a form of cancer in which malignant cells develop in the prostate gland of the male reproductive system. Prostate cancer is the second leading type of cancer in men globally, with an age-standardized incidence rate of 31.1 per 100,000 men . Thus, aEstablished etiological factors of prostate cancer include age, ethnicity, diet, and environmental carcinogen exposure . The int PPARG Pro12Ala polymorphism, NFKB1-94 ins/del polymorphism, NFKBIA-826C/T polymorphism, COX-1 (50C>T) polymorphism, and COX-2 (-1195G>A) polymorphism. The present study aimed to investigate the association of the above polymorphisms with the risk of prostate cancer in the Chinese population.The most common cause of inflammatory response gene dysregulations is mutations in high penetrance genes . DespiteThe study was approved by the Medical Ethics Committee of the Xuanwu Hospital, Beijing, China. All subjects were recruited from Xuanwu Hospital, Beijing, China, between October 2010 and August 2014. A total of 543 males who were diagnosed by qualified physicians as prostate cancer patients based on clinical, pathological, and laboratory tests were recruited into the study. To ensure consistency, only patients with high grade prostate cancer (Gleason score > 7) who had a PSA level of >10 ng/mL were included in the study. At the same period of time, a total of 753 healthy controls were enrolled into this study. Eligibility criteria for control included no individual history of cancer and no known genetic syndromes associated with prostate cancer and a PSA level of <2 ng/mL. Blood samples were collected from the subjects who meet the criteria mentioned above after obtaining written informed consent.μL reaction mixture containing standard PCR buffer, 1.5 mM MgCl2, 0.25 mM dNTP, 1 unit Taq polymerase, and 0.4 μM of each primer. For all polymorphisms, approximately 10% of the samples were chosen at random and sequenced to confirm the genotypes from the amplified PCR product. The laboratory personnel performing the genotyping were blinded to the identity, the case-control status, and source of the samples.Genomic DNA extraction was done by using EasyPure Blood Genomic DNA Kit  after the blood samples were taken from peripheral blood of all subjects. The DNA was used for genotyping of the five SNPs by using the PCR-RFLP method. All PCR were performed in a 25  PPARG Pro12Ala polymorphism were 5′-GCC AAT TCA AGC CCA GTC-3′ and 5′-GAT ATG TTT GCA GAC AGT GTA TCA GTG AAG GAA TCG CTT TCC-3′. The annealing temperature of the reaction was 58°C. The PCR amplification produced a fragment of 270 bp in size, which was then digested using BstUI restriction enzyme. The fragment was cleaved into a 227 bp and a 43 bp fragment for the Ala/Ala genotype while the Pro/Pro genotype was uncut. Apart from that, heterozygous genotype (Pro/Ala) was detected by the presence of all the above bands in agarose gel.The PCR primers used for amplification of the region containing the NFKB1-94 ins/del polymorphism was performed by using 5′-TGG GCA CAA GTC GTT TAT GA-3′ and 5′-CTG GAG CCG GTA GGG AAG-3′ primers, which were annealed to the genomic DNA at 61°C. PflMI restriction enzyme was subsequently used to digest the amplified PCR product of 281 bp in size. For ins/ins genotype, the absence of restriction site resulted in the undigested fragment of 281 bp after the reaction. On the other hand, a 240 bp and a 45 bp fragment were obtained for the del/del genotype, while a 281 bp, a 240 bp, and a 45 bp fragment were found in the ins/del genotype.The PCR for amplification of the region containing the NFKBIA-826C/T polymorphism was genotyped by first amplifying the region of interest by using the following PCR primers: 5′-GGT CCT TAA GGT CCA ATC G-3′ and 5′-GTT GTG GAT ACC TTG CAC TA-3′ at an annealing temperature of 59.5°C. This produced an amplicon of 200 bp in size. Restriction enzyme digestion was performed with BfaI restriction enzyme. The TT genotype was uncut by the enzyme and remained 200 bp, but the CC genotype was cleaved into 180 bp and 20 bp bands. Heterozygotes showed all three bands on agarose gel.TheThe PCR primers used were 5′-GGT GCC CGG TGG GGA ATT TTC-3′ and 5′-GAG GGG AAA GGA GGG GGT TG-3. The annealing temperature used was 60°C. The PCR reaction generated a product of 245 bp. Then, the fragment was digested by using SmuI restriction enzyme. The TT genotype was not cut by the enzyme, hence remaining as a 245 bp band on agarose gels. The CC genotyped, on the other hand, was cleaved into two fragments of 120 bp and 125 bp in size. Heterozygotes showed the presence of all the above bands in agarose gel. COX-2 (1195G>A) polymorphism were 5′-CCC TGA GCA CTA CCC ATG AT-3′ and 5′-GCCTTCATAGGAGATACTGG-3′. The annealing temperature of the reaction was 62°C. The PCR amplification gave a product of 273 bp in size, which was then digested using PvuII restriction enzyme. The AA genotype was undigested, while the PCR product was cleaved into a 220 bp and a 53 bp fragment for the GG genotype. Apart from that, heterozygous genotype (AG) was detected by the presence of all the above bands on the agarose gel.The PCR primers used for amplification of the region containing thet-test. On the other hand, chi square test (χ2 test) was used to determine the significance difference in the smoking status and the distribution of genotypes of the cases and controls. Odds ratios (ORs) and the corresponding 95% confidence intervals (CI) calculated by using logistic regression analysis were used to analyze the association between the polymorphic genotypes and prostate risk, based on heterozygous and homozygous comparison models, with the wild type genotype served as the reference. Odds ratios (ORs) of >1.00 indicated a positive risk association and vice versa. Interaction between the polymorphisms and smoking status, as well as between polymorphisms in related genes , was tested by using the same logistic regression model as above, by using SPSS software. P values of <0.05 were significant. For analyses involving gene-gene and gene-environment interaction, Bonferroni correction was applied for multiple testing, in which case the statistical significance was assumed at P < 0.025.Statistical analyses were carried out using the SPSS software package 18.0 . Mean age for cases and controls was evaluated by using Student's P = 0.29). For both cases and controls, the frequency of never smokers was slightly higher than ever smokers. In cases, 247 (45.49%) were ever smokers and 296 (54.51%) were never smokers, while, in controls, 329 (43.69%) were ever smokers and 424 (56.31%) were never smokers. The difference in the distribution of ever and never smokers among cases and controls was not significant (P = 0.52). NFKB1-94 ins/del polymorphism, the frequency of ins/ins, ins/del, and del/del genotypes was 198, 246, and 99 in cases and 212, 355, and 186 in controls. Significant difference was observed between cases and controls in the distribution of genotypes .The distribution of genotypes of the five polymorphisms in cases and controls is shown in  COX-2 (-1195G>A) polymorphism. Among the cases, 71, 269, and 203 subjects had GG, GA, and AA genotypes, respectively. On the other hand, the distribution of the genotypes among the controls was 158, 378, and 217, respectively .Apart from that, another polymorphism which showed significant difference between cases and controls in the genotypic distribution was theP > 0.05). For NFKBIA-826C/T polymorphism, the distribution of the CC, CT, and TT genotypes was 382, 152, and 9 in cases and 541, 198, and 14 in controls . For COX-1 (50C>T) polymorphism, the CC, CT, and TT genotypes were, respectively, present in 452, 83, and 8 cases as well as 655, 92, and 6 controls . On the other hand, for PPARG Pro12Ala polymorphism, 483, 57, and 3 cases had Pro/Pro, Pro/Ala, and Ala/Ala genotypes, while the distribution of the same genotypes in controls was 676, 73, and 4, respectively .For the three other polymorphisms, no statistically significant difference was observed between cases and controls in the genotypic distribution (P > 0.05).For all polymorphisms, the genotype distribution did not deviate significantly from Hardy Weinberg equilibrium, both among cases and among controls  and homozygous comparison model (del/del versus ins/ins). In heterozygous comparison model, the odds ratio was 0.74 (95% CI = 0.58–0.96), with a P value of 0.02. On the other hand, in homozygous comparison model, the odds ratio was 0.57 . COX-2 (-1195G>A) polymorphism, a significantly increased prostate cancer risk was observed in the heterozygous comparison model (GA versus GG) and homozygous comparison model (AA versus GG). In heterozygous comparison model, the odds ratio was 1.58 (95% CI = 1.15–2.18), with a P value of 0.01. On the other hand, in homozygous comparison model, the odds ratio was 2.08 .For NFKBIA-826C/T polymorphism, statistically not significant risk association was observed. The heterozygous comparison model (CT versus CC) showed an odds ratio of 1.09 (95% CI = 0.85–1.39) and a P value of 0.51. On contrary, the homozygous comparison model (TT versus CC) showed an odds ratio of 0.9104 (95% Cl = 0.39–2.13) and a P value of 0.83.For COX-1 (50C>T) polymorphism and PPARG Pro12Ala polymorphism. Both the polymorphisms showed positive odds ratio values for both heterozygous and homozygous comparison models. For COX-1 (50C>T) polymorphism, the odds ratio obtained in the heterozygous comparison model was 1.31 , while that in the homozygous comparison model was 1.93 . On the other hand, for PPARG Pro12Ala polymorphism, the odds ratio was 1.09 (95% CI = 0.76–1.58) in the heterozygous comparison model (P = 0.63) and 1.05 (95% CI = 0.23–54.71) in the homozygous comparison model (P = 0.95).Similarly, lack of significant association was observed for COX-2 (-1195G>A) polymorphism (P < 0.01), in which a significant risk association was observed between the AA genotype and ever smokers .The interaction between smoking status and the polymorphisms in modulating prostate cancer risk was also formally investigated, and the results are shown in NFKB1 and NFKBIA; COX-2 and COX-1) on the risk of prostate cancer was also examined . For the interactions between COX-2-1195G>A and COX-1 (50C>T) polymorphisms, interaction between COX-2 (-1195AA) ∗ COX-1 (50CT) genotypes conferred a risk increment with OR = 1.83 . However, this was deemed not statistically significant after application of Bonferroni correction, where P < 0.025 was considered as significant. For the interactions between NFKB1-94 ins/del and NFKBIA-826C/T polymorphisms, statistically significant prostate cancer risk association was not observed for any polymorphic combination investigated. No other genotypic interactions yielded a statistically significant association.The effect of interactions between polymorphisms in related genes , and COX-2 (-1195G>A) polymorphisms, with prostate cancer risk in a Chinese population. These polymorphisms have been widely investigated for their associations with other cancers or in other populations, but report is scarce or unavailable for prostate cancer in the Chinese population. This novelty of data represents one of the strengths of the present study. Additionally, our study involved a considerably large number of subjects, which increases the reliability of the results obtained. Apart from that, our study analyzed the combined effect of polymorphisms in a similar pathway , an approach which has been neglected in many previous reports.In this study, we investigated and established the association of five inflammation-related genetic polymorphisms, namely,PPARG gene encodes for peroxisome proliferator-activated receptor-γ (PPAR-γ), which is a ligand-activated transcription factor that is pleiotropic in nature. PPAR-γ is best known for its function in the regulation of adipogenesis and fatty acid metabolism, but it appears that the protein also plays a role in several carcinogenic processes, including apoptosis, cell cycle control, and, notably, inflammation [γ towards its target molecules [ PPARG Pro12Ala polymorphism and prostate cancer risk in our population. We propose that this observation could be due to the fact that although different genotypes of the polymorphism have different binding affinity to its target molecules, the difference is subtle and does not influence the ability of PPAR-γ in activating the transcription of its target molecules. Our result was in agreement with a study from Finland [ammation . The Proolecules . Therefo Finland  and in p Finland , which s NFKB1 and NFKBIA encode for NF-κB1 and its inhibitor, IκBα, respectively. Both NF-κB1 and IκBα are part of the NF-κB pathway, which plays an important role in inflammation and other carcinogenic processes. The -94 ins/del polymorphism of NFKB1 and -826C/T polymorphism of NFKBIA have been shown to affect the transcriptional activity of the respective genes, as the polymorphisms were located in the promoter region of the genes [ NFKB1-94 ins/del polymorphism was associated with decreased risk of prostate cancer in both heterozygous and homozygous comparison models, suggesting that the variant del allele may be linked to a reduced inflammatory status among its carriers and, therefore, a reduced risk. This risk reduction was in agreement with an in vitro assay which showed that the del allele decreased the transcriptional activity of NFKB1, thereby truncating its role in promoting inflammation, which in turn led to the decreased prostate cancer risk [ NFKBIA -826C/T polymorphism, no significant association with prostate cancer risk was observed. Despite many previous studies reported on the association (or lack of association) between the NFKBIA polymorphism and risk of various cancers [On the other hand,he genes , 12. We cer risk . Our fincer risk  in Chinacer risk  in Denma cancers , 23–25,  COX-1 and COX-2 genes, respectively. The COX-1 (50C>T) polymorphism is located within the coding region of the gene and can potentially influence the functional importance of the enzyme produced [ COX-2 (-1195G>A) polymorphism can alter the transcriptional activity of the gene [ COX-1 (50C>T) polymorphism. This indicates that, despite its functional importance, the polymorphism is not sufficiently strong to result in significant risk alteration among its carriers. We also showed the presence of an association between COX-2 (-1195G>A) polymorphism and increased prostate cancer risk in both heterozygous and homozygous comparison models. This suggests that the variant A allele of the COX-2 polymorphism could contribute to an increased prostate cancer risk among its carriers. The observation that the A allele increases prostate cancer risk can be explained by the fact that the A allele creates a c-Myb transcription factor binding site in the gene, resulting in an increased transcription of COX-2 [ COX-1 (50C>T) polymorphism and prostate cancer risk, whereas several previous reports were available for the association between COX-2 (-1195G>A) polymorphism and prostate cancer risk. Our study results were in agreement with Sugie et al. [ COX-2 (-1195G>A) polymorphism and prostate cancer risk.Cyclooxygenases are enzymes which catalyze the conversion of arachidonic acid to prostaglandins, which promote and regulate the process of inflammation. There are two isoforms of cyclooxygenases, namely, the constitutively expressed cyclooxygenase-1 and inducibly expressed cyclooxygenase-2, which are encoded byproduced . On the the gene . Our resof COX-2 . This ine et al.  in a Jape et al. , Cheng ee et al. , and Kope et al.  showed t COX-2 (-1195G>A) polymorphism, a negative risk association for NFKB1-94 ins/del polymorphism, and absence of association for PPARG Pro12Ala polymorphism, NFKBIA-826C/T polymorphism, and COX-1 (50C>T) polymorphism. There are several limitations in this study. First, although the overall sample size was large, the sample size became small when analysis was performed separately between smokers and nonsmokers and when the genotypes were combined. In addition, we only included five polymorphisms which were considered more well established based on our literature review. Therefore, a more extensive examination for a greater number of polymorphisms in a larger sample size is needed and recommended for future work for reconfirmation of our findings.In summary, we have demonstrated the presence of a positive prostate cancer risk association for


LAPTM4B) is a novel cancer-related gene. While recent studies have reported that the LAPTM4B polymorphism increased the susceptibility of several cancers, the results remain inconclusive. Therefore, we performed a meta-analysis to systematically summarize the possible association.Lysosome-associated protein transmembrane-4 beta (LAPTM4B polymorphism and susceptibility to multiple cancers was assessed by odds ratio (OR) and 95% confidence interval (95% CI).The meta-analysis was conducted based on 17 studies in Chinese populations, including 4160 cases and 4148 controls. The relevant studies were searched through electronic databases updated in November 2013. The strength of association between the LAPTM4B polymorphism was significantly associated with overall susceptibility to multiple cancers in all genetic models . Further subgroup analysis revealed a significant association between the LAPTM4B polymorphism and cancer susceptibility in the subgroups stratified by control source, cancer type, histopathologic differentiation, and TNM stage.The meta-analysis results suggested that the LAPTM4B *2 allele was associated with increasing risk of multiple cancers, tumor initiation and development.This meta-analysis indicated that the  LAPTM4B) is a novel cancer-related gene that was cloned from hepatocellular carcinoma (HCC) tissues using fluorescence differential display, rapid amplification of cDNA ends (RACE), and reverse transcription polymerase chain reaction (RT–PCR)[LAPTM4B gene is composed of seven exons and six introns mapped to chromosome 8q22[LAPTM4B by the International Committee[Cancer is one of the leading causes of death in both developed and developing countries, genetic factors contribute significantly to carcinogenesis. Lysosom(RT–PCR),3. Accorommittee.LAPTM4B mRNA and protein were found to be overexpressed in HCC tissues and expressed at fairly low levels in noncancerous or normal liver tissues. In addition, the expression levels were associated with the differentiation status of HCC; it exhibited higher expression in poor-differentiated HCC than in moderate and well-differentiated HCC[LAPTM4B expression was also found in many other malignant tumors, such as breast cancer[LAPTM4B has important roles in many biological processes, including malignant transformation; cell proliferation, invasion, metastasis, and apoptosis; and signal transduction[In previous studies, ated HCC. Recentlt cancer, pancreat cancer, and galt cancer. LAPTM4Bsduction,10.LAPTM4B has two alleles: LAPTM4B *1 (GenBank accession number AY219176) and LAPTM4B *2 (GenBank accession number AY219177)[LAPTM4B allele *1 produces a 317-amino acid protein, while allele *2 was predicted to encode a 370-amino acid protein; there are 53 extra amino acids at the N-terminus of the proteins produced by allele *2[LAPTM4B polymorphism is associated with cancer risk. The LAPTM4B *2 allele is reportedly associated with increased risk of such cancers as liver, gastric, colon, and gallbladder cancer, but not with rectal, oesophageal, or pancreatic carcinoma. Given these controversial results and the limits of a single study with a small sample size, we performed the present meta-analysis of all eligible studies in Chinese populations to investigate the association between the LAPTM4B polymorphism and susceptibility to multiple cancers.Y219177). The difLAPTM4B, Lysosome-associated protein transmembrane-4”, and “polymorphism, mutation, variant”. Reference lists of retrieved articles and reviews were checked for other eligible studies. If the same case series was researched by more than one study, only the one with the largest population was included in our study. If one study investigated multiple cancers, each cancer type was counted as a separate comparison in the subgroup stratified by cancer type.The relevant studies were searched through the PubMed, Embase, Google Scholar, Wanfang, and China National Knowledge Infrastructure (CNKI) electronic databases updated in November 2013. There were no languages, time period, or sample size restrictions. The search key words were “cancer, carcinoma, tumor, neoplasms”, “LAPTM4B polymorphism with cancer susceptibility in a Chinese population; (3) provided sufficient information of genotype distribution in the cases and controls; (4) did not repeat reports in the same population. The criteria for exclusion were (1) case reports, reviews, overlapped data, animal or mechanism studies; (2) no genotype frequency or genotype information provided; (3) not enough information for data extraction.Studies included in the meta-analysis had to meet the following inclusion criteria: (1) original case-control studies; (2) associated the The following data were independently extracted from all eligible studies by two reviewers (Mo and Lu) according to the selection criteria. Disagreements were resolved through group discussion. The data collected from each study were as follows: first author’s name, publication year, ethnicity, cancer type, genotyping method, control source, numbers of cases and controls, genotype distribution in cases and controls, and Hardy–Weinberg Equilibrium (HWE). Eligible studies were divided into hospital-based (HB) and population-based (PB) subgroups, according to the control source. Based on the main cancer type of the included studies, cancer types were classified as gastrointestinal cancer (GIC), including oesophageal, gastric, colon, and rectal cancer; gynaecological cancer (GC), including ovarian, cervical, and endometrial carcinoma; liver cancer, lung cancer, breast cancer (BC), and others, including gallbladder carcinoma, pancreatic cancer, nasopharyngeal carcinoma, and lymphoma.LAPTM4B polymorphism and susceptibility to multiple cancers was assessed by odds ratio (OR) and 95% confidence interval (95% CI) in all genetic models . Subgroup analyses were carried out according to control source, cancer type, tumor grade based on histopathologic differentiation, and tumor stage based on TNM. The stability of the results was assessed using sensitivity analysis by deleting each single study involved in the meta-analysis one at a time to reflect the influence of the individual study to the pooled ORs. Heterogeneity was evaluated by the Chi-square-based Q-statistic and I2 statistic. The DerSimonian–Laird random-effects model was used to assess pooled OR when there was a significant difference in terms of heterogeneity (pQ < 0.1 or I2 ≥ 50%). Otherwise, the Mantel–Haenszel fixed-effects model was used. Potential publication bias was estimated by funnel plots and Egger’s test. An Egger’s test p value <0.05 was considered statistically significant. The HWE was detected by a goodness-of-fit Chi-square test. All p values were two-sided, and p < 0.05 for any test was considered to be statistically significant. This article was prepared based on the PRISMA guidelines, and the checklist was shown in Additional fileAll of the analysis was performed using Stata version 12.0 software . The strength of association between the LAPTM4B gene polymorphism and susceptibility to multiple cancers were identified for detailed evaluation according to our search criteria. After reading the full articles in detail, three studies were excluded because they were case reports[Figure  reports,13 and a reports. Thus, 1 reports-31. Tabl reports,21-29,31 reports,19,20,30 reports. The Zha reports study on reports included reports-18; two  reports; one cer reports; one col reports; one gal reports; one gas reports; one nas reports; one ova reports; one panLAPTM4B polymorphism was significantly associated with overall susceptibility to multiple cancers in the Chinese population in all genetic models  and PB  subgroups. When stratified by cancer type, increased cancer susceptibility was found in all genetic models for liver cancer and GC. There was a significant association with risk of breast cancer in four genetic comparisons, except the homozygous model . In the lung cancer and GIC subgroups, significant associations were only found in the heterozygote model . Moreover, to investigate the relationship between LAPTM4B genotypes and cancer clinical parameters, we analysed the distribution of histopathologic differentiation and classification of TNM and found that the LAPTM4B polymorphism was associated with histopathologic differentiation (heterozygote and dominant models). Significant association was found in the allele, heterozygote, and dominant comparison models in the TNM I–II subgroup and in all genetic models in the TNM III–IV subgroup.The subgroup analysis results are provided in Table In order to test whether the modification of the inclusion criteria of the meta-analysis affected the pooled ORs, we performed a sensitivity analysis by deleting each single study involved in the meta-analysis, one at a time. The pooled ORs were not materially altered, suggesting that our results were statistically robust (data not shown).LAPTM4B polymorphism in the overall population, indicating no obvious publication bias in the meta-analysis.Egger’s test and Begg’s funnel plot were conducted to estimate the possible publication bias of the studies. The Egger’s test results suggested no publication bias for all the genetic models Table . Figure LAPTM4B polymorphism is associated with the risk of several cancers, the results are controversial. Therefore, we carried out this meta-analysis of 17 studies involving 8308 samples, identified through a database search. Overall, we found a significant association between the LAPTM4B polymorphism and susceptibility to multiple cancers in all genetic models. This support from our meta-analysis indicates that the LAPTM4B polymorphism is a possible susceptibility factor for multiple cancers in the Chinese population, especially for liver cancer and GC.Meta-analysis is a valuable tool for estimating the effect of a genetic factor on the risk of a disease when the individual sample sizes are small,33. WhilLAPTM4B is a novel cancer-associated gene belonging to the mammalian LAPTM family and upregulated in a wide variety of solid tumors[LAPTM4B has been shown to promote the transition from G1 to S phase and cell malignant proliferation in LAPTM4B cDNA transfected cells by upregulating the expression of cyclin D1 and E[LAPTM4B may activate the integrin-α6 mediated PI3K-AKT signalling pathway and upregulate some proto-oncogenes such as c-myc, c-jun, and c-fos, which regulate apoptosis and cell proliferation[LAPTM4B and mutated PPRP motif in the N-terminal region of LAPTM4B could attenuate its role in tumorigenesis and metastasis[LAPTM4B*2 could influence the biological activity and function of tumor cells and then induce oncogenic susceptibility[Cancer is a multifactorial disease and its etiology is still unclear. LAPTM4Bd tumors,34. LAPTD1 and E. In addiferation. Moreovetastasis. Previoutibility.LAPTM4B*2 was a risk factor for lung cancer, while Li[LAPTM4B*2 was not closely associated with a susceptibility to lung cancer. Furthermore, the Cheng study indicated that the LAPTM4B polymorphism was associated with the development of colon cancer, but not with rectal or oesophageal cancers[LAPTM4B polymorphism and cancer susceptibility.Previous studies have reported that the increased risk for individuals carrying allele *2 were significantly associated with a susceptibility to liver cancer-20, breawhile Li reported cancers. ConsideLAPTM4B polymorphism and susceptibility to multiple cancers in a large number of Chinese populations. Our results suggested that the LAPTM4B*2 allele was associated with a risk of susceptibility to multiple cancers, which is consistent with the most of the original research. Significant heterogeneity was observed in all of the genetic model comparisons, which might be due to the control sources and mix of cancers. Most of the studies used hospital-based patients as controls, who were not strictly healthy individuals and could not represent the general population. In this meta-analysis, many cancers were only researched once, and some polymorphisms may play different roles in different cancers. After subgroup analyses were conducted to consider the evidence of heterogeneity, some heterogeneity in the control source and cancer type subgroups was found. Therefore, we speculated that environment components and difference of clinicopathological characteristic might contribute to the heterogeneity. In order to assess the influence of the individual data set to the pooled ORs, a sensitivity analysis was carried out by deleting each single study involved in the meta-analysis each time. The results did not alter, suggesting that our meta-analysis results were stable and credible.To the best of our knowledge, this meta-analysis is the first study to investigate the association between the LAPTM4B polymorphism was associated with HB or PB study design, liver cancer, lung cancer, breast cancer, GC, and GIC. Two studies researched the relationship of the LAPTM4B polymorphism with lung and breast cancers, respectively, and they may lack power of statistical analysis. Thirteen studies researched the relationship of LAPTM4B polymorphism with cancer histopathologic differentiation, of which four provided gene frequency data. Eight studies researched the association between LAPTM4B polymorphism and tumor TNM stage, and only three found that LAPTM4B polymorphism was significantly associated with cancer histopathologic differentiation and TNM stage, while the present meta-analysis indicated that LAPTM4B polymorphism was associated with histopathologic differentiation and TNM stage. The small sample sizes in the individual original research studies might account for the false positive correlation. Thus, further studies with larger sample sizes and better design are needed.In further subgroup analyses, we found that the LAPTM4B polymorphism and susceptibility to multiple cancers, and it is statistically more powerful than any single study. Second, all of the eligible studies researched Chinese populations. The participants have the same genetic background, which can reduce the effects of ethnicity on pooled ORs. Third, the distributions of the genotypes in the controls of the included studies met HWE, which increased the statistical power of the meta-analysis. Lastly, we analysed subgroups stratified by control source, cancer types, histopathologic differentiation, and TNM stage. Furthermore, the sensitivity analysis and publication bias analysis indicated that our meta-analysis results were stable and credible. However, there were still some limitations. First, the meta-analysis was based on unadjusted data; the controls were not strictly matched with the cases in some of the original research, which might have caused some bias. Second, there is notable heterogeneity among studies in the overall analysis and in some subgroup analysis, which could lead to some skewed results. Third, the sample size of some of the subgroups was relatively small. In addition, only one study investigated the association in several human cancers. Given these limitations, the present meta-analysis results should be considered with caution. Further studies with larger sample sizes, more cancer types, and other populations are needed to replicate our results.Our meta-analysis has several advantages. First, this is the first systematic review of the association between LAPTM4B*2 allele is associated with elevated susceptibility to multiple cancers. In addition, our results indicate a significant association between the LAPTM4B polymorphism and cancer susceptibility in the subgroups stratified by control source, cancer type, tumor stage based on TNM, and tumor histopathologic differentiation. All of these results indicate that LAPTM4B may be an important biomarker for tumor initiation and development.This meta-analysis indicates that the Written informed consent was obtained from the patient for the publication of this report and any accompanying images.LAPTM4B: Lysosome-associated protein transmembrane-4 beta; HWE: Hardy–Weinberg Equilibrium; HB: Hospital-based; PB: Population-based; GIC: Gastrointestinal cancer; GC: Gynaecological cancer; BC: Breast cancer.All authors declare that they have no competing interests.CM, YL, and XQ conceived and designed the experiments; SL, YD, and JW performed the experiments; LX, TL, and XQ analyzed the data; YH, QP, and JW contributed reagents/materials/analysis tools; and CM and SL wrote the manuscript. All of the authors read and approved the final manuscript.PRISMA Checklist.Click here for file


I2 = 77.4%; n = 14). The RRs were not significant when fats were divided into plant-based fats , animal-based fats , dairy-based fats , saturated FAs , monounsaturated FAs , polyunsaturated FAs , and trans-unsaturated FAs . Similar non-significant results were also observed in most of the subgroup and sensitivity analyses. The findings of this meta-analysis suggest a lack of evidence for associations between dietary fat and FA intakes and EOC risk. Further analyses should be conducted to assess the associations with other types of fat, and the results should be stratified by tumor invasiveness and EOC histology.The associations between dietary fat and fatty acid (FA) intakes and epithelial ovarian cancer (EOC) risk have been inconsistent in previous studies. We conducted a meta-analysis of epidemiological studies to evaluate these associations. We identified relevant studies by searching PubMed, EMBASE, and Web of Science databases. We used random-effects models to estimate summary relative risks (RRs) and 95% confidence intervals (CIs). Overall, the search yielded 20 studies . The summary RR for EOC for the highest versus lowest categories of total dietary fat intake was 1.12 (95%CI= 0.95–1.33;  Ovarian cancer is the most lethal gynecologic cancer worldwide, accounting for approximately 240,000 cases and 150,000 deaths in 2012 .Recent investigations of the pathogenesis of EOC have established that hormones and reproductive status  are the predominant risk and protective factors for this disease , 4, 5. RFigure The characteristics of the 20 selected studies are presented in Table Information collected for the assessment of study quality is presented in Tables I2 = 77.4%) . Non-significant results were also observed regarding intakes of plant-based fat         with significant heterogeneity  Figure  or by thDue to the limitations of the studies included in our meta-analysis, we only completed stratified analyses of intakes of total dietary fat, and saturated, monounsaturated, and polyunsaturated FAs across key study characteristics to explore study heterogeneity Tables  and 5. AI2 = 67.5%, P < 0.001) after omitting the study by La Vecchia et al [I2 = 59.3%, P = 0.003) after omitting the study by Bdoli et al [In a sensitivity analysis of total dietary fat intake and EOC risk, we sequentially removed 1 study at a time and re-analyzed the data. The 14 study-specific RRs ranged from a low of 1.06 , saturated FAs , and animal-based fat  were associated with an increased risk of EOC . HoweverThe present meta-analysis revealed no significant associations between intakes of dietary fat and FAs and EOC risk. Nevertheless, several plausible biological mechanisms have been proposed for the association between fat intake and EOC risk. Dietary fat has been hypothesized to affect ovarian carcinogenesis primarily through hormone-related mechanisms, which were well investigated by the researchers of previous studies. High dietary fat intake may expose the ovarian epithelium to high levels of endogenous estrogens, which may trigger the development of EOC through cell damage and proliferation –9. SeverSeveral recent studies have reported inconsistent associations between dietary fat and FA intakes and EOC risk across different histological subtypes and tumor invasiveness (borderline versus invasive) of EOC. For example, the European Prospective Investigation into Cancer and Nutrition (EPIC) found a borderline significant increased risk for serous EOC with a high intake of polyunsaturated FAs . In contThe difference in point estimates of risk between North American and European populations might be attributed to differences in amounts of FA consumption. Merritt et al  reportedThe present meta-analysis had several strengths. This is the most up-to-date meta-analysis available and it comprised systematic searching, detailed heterogeneity analysis, and study quality evaluations. Additionally, large numbers of EOC cases and non-cases were included, which increased the statistical power to identify clinically meaningful associations. The negative findings were robust in numerous subgroup and sensitivity analyses. However, several potential limitations should be considered. First, although all studies included in this meta-analysis were adjusted for multiple potential confounders, with the exception of 2 studies that provided crude risk estimates calculated from raw data, the possibility of residual confounding by imprecisely or unmeasured factors cannot be excluded. This issue may partly explain the difference between the findings of the Women's Health Initiative Dietary Modification randomized controlled trial and the conclusions of our present meta-analysis, since the former study specifically provided evidence of associations between a low-fat dietary pattern and postmenopausal EOC risk. Therefore, further studies are warranted to rule out residual confounding factors and confirm our findings. Second, the range of dietary fat intake may be underestimated and the magnitude of the associations between intake and risk of cancer may be overestimated because of possible misclassification of fat and FA intakes . HoweverIn summary, the findings of the present meta-analysis, which included 1 pooled analysis, 5 cohort studies, and 14 case-control studies, provide limited evidence of an association between dietary fat intake, including total fat, plant-based fat, animal-based fat, dairy-based fat, and saturated, monounsaturated, polyunsaturated, and trans-unsaturated FAs, and EOC risk. Further prospective studies are needed to confirm the associations between specific types of fat and EOC risk, and the results should be stratified by tumor invasiveness and EOC histology.Two independent investigators (RH and Q-JW) systematically searched PubMed (MEDLINE), EMBASE, and Web of Science databases from each database's inception to the end of May 2015 to identify relevant epidemiological studies. The following keywords were used in the search: (diet OR dietary OR fat OR fatty) AND (ovarian OR ovary) AND (cancer OR tumor OR carcinoma OR neoplasm). Investigators also performed a manual review of references from eligible studies and several review articles , 37. ThiTo be included in this analysis, a study must have (i) been an observational study design; (ii) evaluated the association between dietary fat and FA intakes and EOC risk; and (iii) presented RR, OR, or hazard ratio (HR) estimates with 95% CIs or data necessary for calculation of these risk estimates. Several cohort studies , 40, 41 Studies were excluded if they (i) were randomized controlled trials, retrospective studies, reviews without original data, ecological studies, editorials, or case reports; (ii) reported risk estimates that could not be summarized (such as risk estimates without 95% CIs); and (iii) reported outcomes as EOC mortality or recurrence.Data were extracted by one investigator (Q-JW) using a data extraction form and then entered into a database. An independent investigator (T-TG) checked the data and all differences were resolved by a third investigator (RH). For each included study, we extracted the following information: last name of the first author, publication year, geographic location, number of cases/controls (size of cohort), exposure assessment and categories, and study-specific adjusted estimates with 95% CIs . If there were multiple estimates for the associations, we used the estimate adjusted for the most appropriate confounding variables, as in previous studies –44. In sThe updated NOS , 42–46 rTwo studies , 19 presI2 metric to evaluate the between-study heterogeneity. I2 represents the ratio of between-study variance over the sum of the within-study and between-study variances and ranges from 0% and 100% [versus case-control studies), type of control subjects , geographic location , validated FFQ (yes versus no), number of EOC cases (≥ 450 versus < 450), energy-adjusted model (yes versus no), and adjustment for potential confounders including total energy intake, body mass index, parity, and oral contraceptive use. Heterogeneity between subgroups was evaluated by meta-regression analysis [Similar to our previous studies , 38, we and 100% . We usedand 100% . Pre-speanalysis , 44, 46.P-value of 0.05 was used to determine whether significant publication bias existed. We also conducted sensitivity analyses by deleting each study in turn to reflect the influence of individual data on the overall estimate. All statistical analyses were performed with Stata software .Small study bias, such as publication bias, can reflect genuine heterogeneity, chance, or other reasons for differences between small and large studies; we evaluated study bias with Egger's regression asymmetry test . A P-val


Since the late 1960s, the average global supply of fat has increased by 20 g per capita per day. While fat intake has been considered a potential risk factor for prostate cancer (Pca), the hypothesis from previous epidemiologic studies remained equivocal.Relevant cohort studies were identified through a literature search in PubMed, ScienceDirect and Wiley Online Library up to March 1, 2015. A systematic review and dose-response meta-analysis were used to assess the relationship between fat intake and the risk for Pca.We identified 14 cohort studies, which included 37,349 cases and a total of 751,030 participants. We found no evidence of a non-linear association between fat intake and the risk for Pca. Overall, the summarized relative risks for every 28.35 g increment a day was 0.99  for total fat intake, 1.00  for saturated fat, 0.99  for polyunsaturated fat, and 1.00  for monounsaturated fat. Additionally, there was no link to the risk for advanced stage Pca regarding total fat intake , saturated fat , polyunsaturated fat , or monounsaturated fat . Subgroup and sensitively analyses showed consistent results.Little evidence from published cohort studies supports the statement that total fat, saturated fat or unsaturated fat intake increases the risk for Pca or advanced stage Pca. Prostate cancer (Pca) is the second leading cause of cancer death among American males  and has The World Health Organization reports that since the late 1960s, the average global supply of fat has increased by 20 g per capita per day . In manyIn several previous reviews and meta-analyses , 13, 14,We conducted our meta-analysis following the preferred reporting method for systematic reviews and meta-analyses (PRISMA) statement  S1 PRIS.Eligible cohort studies were identified by searching PubMed, ScienceDirect, and Willey Online Library published up to March 1, 2015. Two reviewers (T.Z. Liu and Z.Y. Gao) independently searched each database and any disagreements were resolved by a methodologist (X.T. Zeng) for a final decision. A Kappa statistical test was used for measuring agreement . The folBecause case-control studies may introduce considerable bias, particularly recall bias, only cohort or case-cohort studies were included in our meta-analysis , 18, 19.From all included studies two experienced reviewers (S. Li and F.F. Han) extracted the first author’s name, publication year, country, study type, follow-up, age distribution, types of fat number of cases or person-years, serving size, adjusted or crude relative risk (RR) with 95% confidence intervals (CI), adjusted variables, and the degree of Pca using a standardized data collection sheet. When different models were used to adjust for confounders, we extracted the RR that controlled for the most factors. A third investigator (Z.Y. Gao) checked the data and corrected potential errors.For studies that measured fat intake with energy percentage, the data was converted to grams by multiplying by the mean daily total energy intake and then dividing by 9 . If studies did not report the mean daily energy intake, we assumed it to be 2,418 Kcal, which is the age–specific energy value for 50–71 year old males according to the National Cancer Institute . This daRelative risk (RR) was used to measure the risk. Odds ratios (OR) and hazard ratios (HR) were roughly regarded as relative risk (RR) . MissingThe dose-response meta-analysis was conducted in two steps. First, the generalized least-squares method estimated the coefficient per unit increment of exposure within each study. Second, the regression coefficients were combined in a random-effect model with the weight calculated by inverse variance , 24. AllWe used the mean value of the lower and upper boundaries of each category as the assigned dose. For open-ended lower categories, the assigned dose was calculated by dividing the cut-off point by 1.2. For open-ended upper categories, the cut-off point was multiplied by 1.2 . The non2 statistic assessed heterogeneity, and subgroup and sensitivity analyses evaluated whether the results were consistent. A random-effects meta-regression was used to assess which covariates in the subgroup analysis influenced the intervention effect [The Egger’s test was used to determine publication bias, the In effect . All theThere were 204 search results in PubMed, 154 in ScienceDirect and 262 in Willey Online Library. After eliminating cohort studies with duplicate and unrelated results, including one study  that didAmong the 14 studies, there were a total of 751,030 participants of which 37,349 cases developed Pca. The mean age of Pca cases was about 60.60 years and the mean follow-up period was 9.2 years. Four studies , 37 measThe Newcastle-Ottawa Scale , 43, whi2 = 5.0%, P = 0.34) (Thirteen studies –41 inves = 0.34) .2 = 14.3%) for every 28.35 g (1 ounce) increment of saturated fat consumed per day  for total unsaturated fat intake, 0.99  for polyunsaturated fat, and 1.00  for monounsaturated fat (As to a linear association (every 28.35 g increment per day), the combined RR was 0.99  for total fat, 0.96  for saturated fat, 0.96  for polyunsaturated fat, and 0.96  for monounsaturated fat intake.Seven studies , 34, 37 P-value was 0.32 for total fat, 0.71 for saturated fat, 0.91 for polyunsaturated fat, and 0.65 for monounsaturated fat consumption.We conducted a subgroup analysis using primary measurement units (e.g. grams or energy), study area, and adjustment status (adjusted/not) for BMI for potential divergences within subgroups, which showed no substantial change . The mulSensitivity analysis using the random-effect model showed robust results after omitting one study at a time from each analysis. It also showed consistent results when two studies , 33 withThe Egger’s test did not detect obvious publication bias in our meta-analysis of studies that reported total fat (P = 0.93), monounsaturated fat (P = 0.16), and polyunsaturated fat consumption (P = 0.92). However, obvious asymmetry was observed in saturated fat (P = 0.01). An adjusted meta-analysis using the trim and fill method showed consistent results in both fixed-  and random-effects .The present meta-analysis confirmed no obvious associations between total, saturated or unsaturated fat consumption and the risk for Pca. Our further analysis also confirmed a lack of association between fat intake and advanced or high grade Pca. Although obvious asymmetry was observed in the Egger’s test of saturated fat intake, the trim and fill method showed consistent results and suggest that asymmetry may be not be caused by publication bias.Obesity is often linked to high fat consumption  while BMThe sensitivity analysis showed robust results in our meta-analysis. Two studies , 33 accoA previous meta-analysis , of 29 oThere are a few known risks and benefits to consuming dietary fat. Bioactive components in dietary fats such as N-3 polyunsaturated fat acid (n-3 PUFA), may protect against prostate cancer and other types of cancer  by alterA study by Mills et al  of 180 cPhysical activity status may influence our results since regular physical activity is considered to protect against Pca . Two stuOther confounding factors such as age, energy, family history, fruit and vegetable intake, and serum fatty acid levels may have influenced our results. However, sensitivity analysis did not reveal substantial change, which suggests that the influences mentioned above may have little impact on our results.To ensure that our results are reliable, we used a dose-response meta-analysis to evaluate high quality cohort studies and to find potential non-linear or linear relationships between fat intake and the risk for Pca. We used subgroup and sensitivity analyses to distinguish the effect among the subgroups and generated consistent results. All of this makes our results more reliable.There were a few limitations in our meta-analysis as well. First, the limited number of studies included and the considerable amount of heterogeneity detected in our analysis of advanced or high grade Pca and fat consumption may have influenced the accuracy of our results. Second, all studies were conducted in American or European countries. Thus, a selection bias was introduced making this meta-analysis applicable to Americans and Europeans only. Third, we have limit data on addressing the above potential bias in our study that may also influence our results.Current published cohort studies suggest no association between total fat, saturated fat, or unsaturated fat intake and the risk for Pca. More studies on the association between fat intake and high grade or advanced stage Pca are needed.S1 FigThe P value for non-linear test was 0.97. The points assigned to 4.17 g (reference dose), 10.2 g, 15.31 g, 19.88 g, and 25.47 g, respectively.(TIF)Click here for additional data file.S2 FigThe P value for non-linear test was 0.54. The points assigned to 15.74 g (reference dose), 25.58 g, 35.73 g, and 45.1g, respectively.(TIF)Click here for additional data file.S1 PRISMA Checklist(DOC)Click here for additional data file.S1 Table(XLSX)Click here for additional data file.S2 Table(XLSX)Click here for additional data file.


The PlA1/A2 polymorphism of glycoprotein IIIa (GPIIIa) has been reported to be associated with risk of stroke in some studies, although other studies suggest no such association. This meta-analysis and systematic review was conducted to investigate the hypothesis that carriage of the PlA2 allele is a risk factor for stroke.Electronic databases (MEDLINE and EMBASE) were searched for all articles evaluating carriage of the PlA2 allele and the incidence of stroke. Pooled odds ratios (ORs) were calculated using fixed-effect and random-effect models.n = 11,873; 95% CI = 1.03–1.22; p = 0.011) was observed for the incidence of stroke in adults, with subgroup analyses identifying the association driven by stroke of an ischaemic  but not haemorrhagic aetiology . This association with ischaemic stroke was strongest in individuals homozygous for the PlA2 allele compared to those homozygous for wild-type PlA1 . Subgroup analysis of ischaemic stroke subtypes revealed an increased association with stroke of cardioembolic  and large vessel  aetiology, but not those of small vessel origin . Egger's regression test suggested a low probability of publication bias for all analyses (p>0.05).A total of 35 articles were eligible for inclusion, of which 25 studies were suitable for statistical analysis. For carriage of the PlA2 allele, OR 1.12 (The totality of published data supports the hypothesis that carriage of the PlA2 polymorphism of GPIIIa is a risk factor for ischaemic strokes, and specifically those of cardioembolic and large vessel origin. The platelet fibrinogen receptor is integral to primary haemostasis, since it regulates platelet aggregation and the formation of stable thrombus following vascular injury n = 9,547; odds ratio (OR) 1.205, 95% confidence interval (CI) 1.067–1.360; p = 0.003) In 1996, carriage of the PlA2 allele of GPIIIa was reported to be associated with myocardial infarction (MI) when compared to subjects homozygous for the PlA1 wild-type MI and ischaemic stroke are modulated by the same cardiovascular risk factors, although differing arterial specificity for these risk factors underlies subtle differences in their respective pathophysiology The PlA2 polymorphisms of GPIIIa may transcend these pathophysiological discrepancies as, lying adjacent to the ligand binding site of the fibrinogen receptor, it is positioned to modulate the final common pathway of platelet-mediated thrombosis. Indeed, a meta-analysis published in 2010 considering 187 candidate genes for ischaemic stroke identified 14 studies examining the PlA2 polymorphism, and suggested the presence of a possible association Here we present the most comprehensive meta-analysis to date of the association between carriage of the PlA2 allele and ischaemic stroke, with inclusion of 25 studies and 11,873 subjects. Subgroup analyses based on subject demographics and ischaemic stroke subtypes are presented. Additionally, we report the first meta-analysis examining the possible association of the PlA1/A2 polymorphism with haemorrhagic stroke in 1,621 subjects, to further explore the relationship between this polymorphism and cerebrovascular events.th July 2013 for all articles evaluating carriage of the PlA2 polymorphism of GPIIIa and the incidence of stroke. The medical subject headings (MeSH) terms used in the primary literature search were ‘stroke’, ‘cerebrovascular’, ‘subarachnoid’, ‘subdural’, and ‘extradural’ in combination with ‘glycoprotein IIIa’, ‘integrin beta3’ and their associated synonyms. No language restrictions were in place during the primary search. A secondary search of the literature was performed by hand, based on all potentially relevant articles identified from the primary search.Electronic databases (MEDLINE and EMBASE) were searched up until 30Articles were considered suitable for inclusion in the systematic review if they studied association between carriage of the PlA2 polymorphism and risk of stroke. Studies performed wholly in children (<18 years) were excluded, although studies that considered ‘young adults’ (often defined as 15–44 years) were included. Non-English language articles were included if sufficient information could be extracted from the abstract.In the case of studies suitable for statistical analysis, the following data were extracted where available: the number of cases/controls for each genotype , stroke subtype and population demographics . For duplicated studies, the most complete dataset was utilised. Where more than one control group was available, those with cardiovascular risk factors were selected in preference to healthy subjects. Where genotype information was reported for >1 subpopulation as defined by geographic region or ethnic origin, each subpopulation was considered separately in the analyses Data were analysed using Comprehensive Meta-analysis version 2 software . Pooled ORs were calculated using fixed- and random-effects models, along with the 95% CIs to measure the strength of association. Fixed-effects summary ORs were calculated using the Mantel-Haenszel method 2 was also calculated for each analysis, where ≥50% may represent substantial heterogeneity Tests for heterogeneity were performed for each analysis, with significance set at p<0.05 n = 1) n = 1) n = 4) n = 2) n = 4) The primary search yielded 2,644 unique articles, of which 43 were identified as being potentially relevant following review of title and abstract The secondary search yielded an additional three articles n = 11,873; 95% CI = 1.03–1.22; p = 0.011). Significant heterogeneity was observed , and the significance of association with the PlA2 allele was lost with application of the random-effects model .Carriage of the PlA2 allele was significantly associated with total stroke, with an OR 1.12  . Analysis using the random-effect model revealed a significant association of similar magnitude . Subgroup analysis of subjects homozygous for the PlA2 polymorphism demonstrated a greatly increased degree of association   , with sip<0.001) .The ten studies assessing carriage of the PlA2 allele did not include suitable data for statistical analysis and are summarised in n = 1,621; OR = 0.90; 95% CI 0.71–1.14; p = 0.398)    or those= 0.816) .n = 1,394; OR 2.76. 95% CI 1.34–2.31; p<0.001) and cardioembolic (CE) disease , but not with stroke secondary to small vessel disease (SVD)   and CE subgroups .The magnitude of the association observed between carriage of the PlA2 allele and ischaemic stroke was not independent of aetiology of the vascular event . Signifi= 0.950) . Substann = 3,843; OR 1.22, 95% CI 1.05–1.42; p = 0.009).Data were available to calculate pooled ORs based on subject age, ethnicity and sex . No subgPublication bias was assessed by construction of funnel plots and calculation of Egger's regression intercept. All analyses demonstrated a low probability of publication bias (p<0.05). Furthermore, subgroup analyses based on study size did not suggest increased publication of positive results by smaller studies. For studies with <250 cases, OR was 1.15 , and for studies with ≥250 cases, OR was 1.16  .The results of this study demonstrate that carriage of the PlA2 allele of GPIIIa is a strong risk factor for ischaemic stroke. The validity of this association is strengthened by the finding that subjects homozygous for the mutant allele have almost five times the risk of stroke compared to those subjects who are heterozygous.An age-skewed risk profile caused by a relative decrease in the influence of genetic factors with increasing age might have been expected, given the results of our recent meta-analysis of the association between PlA2 carriage and risk of MI, which demonstrated a stronger association in younger patients Subgroup analyses were performed for ethnicity, sex and study size, the results for sex and study size being consistent with the primary analysis suggesting that these factors are not major modulators of the increased risk associated with the polymorphism. Caucasian ethnicity produced a significantly higher association, suggesting that there may be ethnic differences in the impact of carriage of the PlA2 allele. Analysis based on ethnicity was clearly limited by the underlying low prevalence of the PlA2 allele in non-Caucasian populations. The prevalence of the allele in Asians is low, and this was emphasised in three studies of Chinese populations that identified no carriers of the mutant allele across their 554 cases Subgroup analyses based on ischaemic stroke subtype identified an increased association between carriage of the PlA2 allele and stroke of CE or LVD aetiology. These findings are based on three studies per subtype and contain low numbers of cases (520 and 452 respectively). However, much published data suggest that platelet activation modulates CE and LVD stroke rather than SVD. In a prospective study of 54 cases of non-CE stroke, those with LVD were found to have significantly higher platelet expression of CD63 during the acute phase, and significantly higher CD62P (P-selectin) expression up to 30 days post event, thus indicating higher levels of platelet activation in vitro investigations into the effect of the proline substitution on ligand binding have observed increased adhesion to fibrinogen and von Willebrand factor (vWF) in cell culture under conditions of shear stress max) or dissociation constant (Kd) were observed A further divergence in the aetiology of ischaemic stroke between LVD and SVD may arise because of differences in wall shear stress in the respective vessels. The mechanism by which platelets aggregate is dependent on shear stress, with transient aggregation occurring in the absence of activation at high shear rates Ex vivo studies have provided conflicting data on the effect that carriage of the PlA2 allele has on platelet fibrinogen binding and on platelet aggregation, in both healthy and ischaemic stroke cohorts There is a suggestion, from the results of this study, of an apparent decrease in the risk of haemorrhagic stroke in subjects carrying the PlA2 allele. This however was non-significant, but it should be noted that this analysis was based on a relatively small pool of 2,470 subjects, and merits further investigation in future studies. If confirmed, this would give further insight into the mechanism by which the polymorphism affects platelet function. As discussed above, the amino acid substitution lies adjacent to the ligand binding site of the fibrinogen receptor and so may modulate normal platelet function; however, whether this is truly so remains contentious. in vivo studies appear to be more consistent, with carriage of the PlA2 allele resulting in a reduction in bleeding time both pre- and post-aspirin therapy in healthy subjects Observations from ex vivo findings may possibly be due to the impact of the polymorphism being dependent on interaction of platelets with the vasculature, and this hypothesis warrants further investigation in future in vitro studies.The totality of these experimental findings, when viewed in the context of the clinical data reported in the present study, suggests that carriage of the PlA2 allele does indeed result in modulation of platelet function leading to increased activation and aggregation. The clinical manifestation of this modulation is an increased propensity for thrombus formation. The discrepancy with There are a number of limitations to the study presented here that are inherent to the data available. Due to the increase in the prevalence of traditional cardiovascular risk factors with age, it is preferable to include data that are adjusted for these factors as the contribution of a single gene polymorphism to a multifactorial, polygenic process is likely to be subtle The majority of strokes occur out-of-hospital with many patients not surviving the index event We have demonstrated that carriage of the PlA2 polymorphism of GPIIIa is associated with an increased risk of ischaemic stroke in adults, and this risk is higher in homozygous subjects and those with strokes of CE and LVD aetiology. This risk is likely to be modulated by increased platelet activity that, in conjunction with interaction with the vascular wall, results in increased thrombus formation. The magnitude of the increased risk in PlA2 carriers is similar to that seen in MI, and suggests that PlA1/A2 genotyping may add usefully to risk stratification for patients at risk of stroke.Checklist S1PRISMA Checklist.(DOC)Click here for additional data file.


P = 0.118 for heterogeneity, I2 = 37.6 %). In addition, when subgroup analyses by study quality and number of cases, more statistically significant associations were observed in studies of high quality  and large sample size . There was no evidence of significant publication bias with Begg's test (P = 0.602) or with Egger's test (P = 0.350). Overall, this study indicates that use of RAS inhibitors may be associated with a decreased risk of prostate cancer. Large-scale well designed studies are needed to further explore this association.Emerging evidence suggests that renin-angiotensin system (RAS) may act as a molecular and therapeutic target for treating site-specific cancers, including prostate cancer. However, previous observational studies regarding the association between RAS inhibitors and prostate cancer risk have reported inconsistent results. We examined this association by performing a systematic review and meta-analysis. A total of 20,267 patients from nine cohort studies were enrolled. Compared with non-users of RAS inhibitors, individuals using RAS inhibitors had a reduced risk of prostate cancer , without statistically significant heterogeneity among studies ( Prostate cancer is the second most commonly diagnosed cancer in men worldwide, with 1,111,700 new cases and 307,500 deaths estimated to have occurred in 2012  and crucRenin-angiotensin system (RAS) inhibitors, such as angiotensin-converting enzyme inhibitors (ACEIs) and angiotensin II receptor blockers (ARBs), are commonly used in the treatment of hypertension across the world. Beyond cardiovascular effects, there is some evidence that RAS inhibitors might reduce risk of site-specific cancers , Europe (n = 4), and Asia (n = 1). All of the included studies were published between 2001 and 2013, including a total of 20,267 cases. Information on exposure (RAS inhibitors) and outcome (prostate cancer) was mainly obtained by medical records. Four studies used hazard ratio (HR), two used RR, two used odds ratio (OR), and one used standardized incidence ratio (SIR). The study quality scores, assessed by the NOS, ranged from 5 to 8 (with a mean of 7). Table The detailed steps of our literature search are presented in Figure Asia n = . All of P = 0.012) was observed among individuals using RAS inhibitors. There was moderate but not statistically significant heterogeneity among studies .The multivariable-adjusted RRs for each study and for the combination of all included studies are shown in Figure Next, we carried out subgroup analyses by study design, geographical region, study quality, and number of cases Figure . When st2 index to assess heterogeneity in this meta-analysis. As shown in Figure P = 0.118 for heterogeneity, I2 = 37.6 %). Then we performed Galbraith plot analysis and found that studies by Friis et al. [P = 0.606, I2 = 0.0 %) and the combined RR remained statistically significant  or with Egger's test .There was no evidence of significant publication bias with Begg's test  databases. Considering some studies that assessed the relationship between RAS inhibitors and overall cancer risk also reported site-specific cancer risk, we broadened the search by using a loose search algorithm (see i) the exposure of interest was RAS inhibitors; (ii) the outcome of interest was prostate cancer; (iii) study design was cohort or nested case-control; and (iv) the effect size estimates with their corresponding 95 % confidence intervals (CIs) were provided .The studies included in this meta-analysis met all of the following criteria: . NOS is an eight-item instrument that allows for the assessment of selection (four items), comparability (one item), and outcome (three items). Each item corresponds to one point, except for comparability (two points). Thus the range of possible scores is 0-9 points. Studies are assigned as high quality if the score is 7-9 points.Two authors (Y.Q.M and X.X.) evaluated the quality of each study by using the Newcastle-Ottawa Scale (NOS)  using a predefined data collection form. Discrepancies were resolved by consensus or consulting a third reviewer.Relative risks (RRs) and their 95 % CIs were used to calculate and assess the strength of the association between use of RAS inhibitors and prostate cancer risk. A random-effects model proposed by DerSimonian and Laird , which t2 index [2 was used to quantify the impact of heterogeneity . Furthermore, the Galbraith plot [The heterogeneity of RRs across the studies was tested by the Cochran Q and the I2 index . The levith plot  was usedSensitivity analysis was performed whereby each study was removed in turn and the pooled estimate was recalculated to determine the influence of each study. Cumulative meta-analysis was also carried out by sorting the studies based on publication year or from most precise to least precise.P values .Potential publication bias was assessed using Begg's test (rank correlation method)  and Egge


The etiological factor for intrahepatic cholangiocarcinoma (ICC) is not clear. Although it has been widely accepted that intrahepatic biliary tree stone is associated with increased risk of ICC, the role of extrahepatic biliary tree stone in the incidence of ICC is controversial. In the present study we aim to evaluate the association between pre-existing choledocholithiasis and cholecystolithiasis and the risk of ICC.PubMed, Embase, and Web of Science were searched to identify cohort and case–control studies on the association between choledocholithiasis or cholecystolithiasis and the risk of ICC. Studies that met the inclusion criteria were subjected to a meta-analysis performed with Stata statistical software. Either a fixed or random effect model was used, depending on the heterogeneity within the studies. Egger’s test was performed to assess publication bias.I2 = 78.5 %).Seven case–control studies met our inclusion criteria. Of the 123,771 participants, 4763 (3.85 %) were patients with ICC, and 119,008 were tumor-free controls. The presence of pre-existing bile duct stones  was associated with the risk of ICC . Even the presence of choledocholithiasis alone (in the absence of hepatolithiasis) was associated with a high risk of ICC . Cholecystolithiasis may possibly contributed to the incidence of ICC , with large heterogeneity within studies (Bile duct stones including choledocholithiasis are important risk factors for ICC. Careful surveillance of patients with extrahepatic biliary tree stone should be considered.The online version of this article (doi:10.1186/s12885-015-1870-0) contains supplementary material, which is available to authorized users. Biliary tract neoplasms are classified as intrahepatic cholangiocarcinoma (ICC), perihilar cholangiocarcinoma, or extrahepatic cholangiocarcinoma depending on the tumor location within the biliary tree . ICC, whPubMed, Embase, and Web of Science were searched using the following key words: ‘Cholelithiasis’ or ‘Choledocholithiasis’ or ‘Cholecystolithiasis’; ‘Intrahepatic cholangiocarcinoma’ or ‘cholangiocarcinoma’ or ‘bile duct neoplasms’; and ‘Risk factors’ through December 2014. No limitations were set for the language or the year of publication. The reference lists of the retrieved articles were manually searched so that no possibly useful information was missed.The inclusion criteria were as follows: 1) cohort or case–control studies of the correlation between ICC and pre-existing choledocholithiasis (with or without concurrent hepatolithiasis), or cholecystolithiasis independently; (2) studies in which the primary outcome was the occurrence of ICC; (3) studies in which the exposure of interest was the presence of either pre-existing choledocholithiasis (with or without concurrent hepatolithiasis) or cholecystolithiasis; and (4) studies in which estimates of relative risk  with their 95 % confidence intervals (CIs) were available. Studies that enrolled patients with concurrent bile duct stones and cholecystolithiasis were excluded. The flow chart for selection of the studies is shown in Fig.  cohort oTwo reviewers  independently extracted the data from the selected studies using a specially designed form. The following information was required for our study: name of first author, publication year, country or region, study design, number of cases (incidence of ICC in cohort studies), number of controls or cohort size, matched factors and confounders of each study. The validated Newcastle–Ottawa scale was used to assess the methodological quality of case–control and cohort studies . Any disχ2 test. The Cohen’s Kappa was used to assess the inter-rater reliability for inclusion decision. Relative risks  with their corresponding 95 % CIs were used to assess the association between the risk of the development of ICC and pre-existing biliary stone disease. χ2 and I2 tests were used to assess between-study heterogeneity; I2 values ≥ 25, 50 and 75 % were considered to indicate mild, moderate and high heterogeneity. Either a fixed or random effect model (Inverse Variance method) was used, depending on the between-study heterogeneity. Subgroup analysis was performed to identify confounding factors that could possibly contribute to between-study heterogeneity. Publication bias and other biases were assessed by means of Egger’s test. Trim and fill tests combined with conversion between different effect models were performed in sensitivity analysis. P ≤ 0.05 was considered to indicate statistical significance.Meta-analysis was performed with Stata statistical software . Dichotomous variables were expressed as relative frequencies and were compared by means of the As shown in Fig. Nordenstedt et al. conducted a cohort study on the association between cholecystolithiasis and the risk of ICC in a Swedish population. However, patients who had both cholecystolithiasis and bile duct stones were not eliminated, so this study was excluded from our analysis . Wu et aI2 = 49.8 %, p = 0.077), and a random effect model was used. Subgroup analysis based on different regions, study designs and NOS scores, which are possible confounding factors was performed and shown in Table As shown in Fig. I2 = 78.5 %, p = 0.000). Subgroup analysis based on different regions, study designs and NOS scores was performed and shown in Table As shown in Fig. t = −2.37, p = 0.077). However, biases existed in the meta-analysis of cholecystolithiasis , which could be due to high heterogeneity within studies. A sensitivity analysis was performed. Trim and fill analysis showed that the outcomes were not changed without trimming performed, and the outcomes still showed statistical significance when a fixed effect model was used.Egger’s test showed no evidence of publication bias for the meta-analysis of bile duct stones . There is a strong correlation between hepatolithiasis and ICC as confirmed by literature, which is in accordance with our findings . Our metStatistically significant heterogeneity existed within the studies included in the meta-analysis for cholecystolithiasis, which may be attributable to differences in regions (eastern versus western countries), study designs , and Newcastle–Ottawa scale scores , as shown in Table Our meta-analysis does have some limitations. First, the evidence levels of the included case–control studies were relatively low, and there were no qualified cohort studies. Cholelithiasis is usually accompanied by other metabolic syndromes, which are also risk factors for the development of ICC –25. To rAs far as we know, this is the first systematic review and meta-analysis of studies evaluating the association between the risk of developing ICC and pre-existing choledocholithiasis and cholecystolithiasis, and our study may be of value for clinical practice. The prognosis of ICC is extremely poor, so early diagnosis and timely treatment of this highly malignant disease are important. Our evidence-based study showed that patients with a history of cholelithiasis, especially choledocholithiasis, are at high risk of developing ICC. Therefore, routine follow-up for these patients is critical for the early diagnosis of ICC. Early diagnosis and timely treatment can be expected to lead to better outcomes for ICC patients.Bile duct stones were found to be important risk factors for the development of ICC. Even in the absence of hepatolithiasis, choledocholithiasis was associated with a high risk of ICC.


G = 1.41, 95%CI:1.21-1.65). For MMP2-1306 C/T and -735 C/T, based on the model-free approach, a significantly reduced risk was found in Asians. For MMP9-1562 C/T, a significantly increased risk was found among Asians with model-free approach. For MMP13-77A/G, there was no association between this polymorphism and lung cancer risk in the recessive model and with the model-free approach. Therefore, this meta-analysis suggests that the MMP1-1607 1G/2G, MMP2-1306 C/T, MMP2-735 C/T, MMP9 -1562 C/T polymorphisms were risk factors for lung cancer among Asians, while MMP13 -77A/G polymorphism was not associated with lung cancer risk.Many studies have reported the association between the matrix metalloproteinase (MMP) polymorphisms and lung cancer susceptibility, but the results were inconclusive. We conducted a meta-analysis, using a comprehensive strategy based on the logistic regression and a model-free approach, to derive a more precise estimation of the relationship between MMP1, MMP2, MMP9 and MMP13 polymorphisms with lung cancer risk. A total of 22 case-control studies including 8202 cases and 7578 controls were included in this meta-analysis. For MMP1-1607 1G/2G, increased lung cancer risk was found among Asians in additive model and with model-free approach(OR Lung cancer is the leading cause of cancer-related death worldwide and responsible for approximately 1.3 million deaths each year3The matrix metalloproteases (MMPs) are zinc-dependent endopeptidases that degrade the extracellular matrix collagens and belong to a larger family of proteases known as the metzincin superfamily. 556710Polymorphisms in the regulatory regions of MMPs have been associated with changes in the expression level of these genes in different human cancer12131516171819202122232425262728293032We searched for relevant studies up to May 2014 through the PubMed, Embase, Wanfang (http://www.wanfangdata.com.cn), China National Knowledge Infrastructure Platform  database with the following terms and their combinations: “lung cancer/carcinoma”, “polymorphism/variant”, and “metalloproteinase/MMP”. We tried to identify potentially relevant studies from the whole reference lists by orderly reviewing title, abstract and full text.The inclusion criteria were as follows: a) case-control studies focused on the association of MMP1, MMP2, MMP9 or MMP13 polymorphism and lung cancer; b) genotype and allele data available. Studies were excluded for following reasons: a) unpublished papers, reviews and duplication of publications; b) data unavailable for calculating genotype or allele frequencies; c) no control population. Additionally, investigations of departure from Hardy-Weinberg equilibrium (HWE) was excluded from the final analysis. If more than one article was published using the same case series, we selected the study with the largest sample size.All the available data were extracted from each study by two investigators (H X L and X Y L) independently according to the inclusion criteria listed above. For each study, we recorded the first author, year of publication, country of origin, ethnicity, the method of genotyping, the number of cases and controls and genotype distributions in cases and controls.P > 0.05) using gene frequencies of the healthy individuals. Metagen (http://bioinformatics.biol.uoa.gr/~pbagos/metagen/) was used by selecting the genetic model. Two parameters, θ2 and θ3, were calculated using the formula: log it (πij) = αi + θ2zi2 + θ3zi3 and OR AB/AA = exp (θ2), OR BB/AA = exp (θ3); where αi is the indicator of study-specific fixed-effect; θ2 and θ3 are dummy variables of genotypes AB and BB. The appropriate genetic model was identified using the following criteria:(i) No association: θ2 = θ3; (ii) Dominant model: θ2 = θ3 > 0; (iii) Recessive model: θ2 = 0 and θ3 > 0; (iv) Additive model: 2θ2 = θ3; (v) Co-dominant model: θ3 > θ2 > 0; (vi) Complete overdominant model: θ2 > 0 and θ3 = 0. Finally, once the most appropriate genetic model was identified, the pooled OR with corresponding 95% confidence interval (95% CI) was estimated in logistic regression model. Additionally, Zintzaras reported a novel method to calculate the generalized odds ratio (ORG) based on a genetic model-free approach, which may overcome the short-comings of multiple model testing or erroneous model specificationG calculations were also performed.Hardy-Weinberg equilibrium was examined by chi-square goodness-of-fit test (I2 statistic. Both fixed effects (Mantel-Haenszel) and random effects (Der Simonian and Laird) models were used to combine the data. Relative influence of each study on the pooled estimate was assessed by omitting one study at a time for sensitivity analysis. Publication bias was evaluated by visual inspection of symmetry of Begg’s funnel plot and assessment of Egger’s test . Statistical analyses were done in ORGGASMA, metan and metagen in STATA software, version 11.0 , and all tests were two-sided.The heterogeneity of the studies was assessed using the Cochran’s Q test (considered significant for P < 0.10) and was quantified by the There were 330 papers relevant to the search words. The flow chart of selection of studies and reasons for exclusion is presented in There was a variation in the 2G allele frequency of the MMP1-1607 1G/2G polymorphism among the controls across different ethnicities, ranging from 0.46 to 0.71. For Asian controls, the 2G allele frequency was 0.56, which was slightly higher than that in Caucasian controls  and 1.16(95%CI = 1.02-1.33), respectively, suggesting an additive model was assessed using traditional method. Overall, no significant association with lung cancer risk was detected for MMP1 -1607 1G/2G polymorphism in additive model and heterogeneity between studies was observed in the overall comparison. In subgroup analysis based on ethnicity, however, the heterogeneity disappeared and a significantly increased risk was found in Asians . No significant association was found in subgroup analyses based on the source of control and sample size  . Based ople size .CT/CC and ORTT/CC were 0.54(95%CI = 0.47-0.63) and 0.53(95%CI = 0.33-0.85), respectively, suggesting no appropriate genetic model was assessed using traditional method. Based on the model-free approach, significant result was found in the overall comparison  and among Asians , but not among Caucasians  . Stratif74-1.59) . When st74-1.59) . No signCT/CC and ORTT/CC were 0.70(95%CI = 0.59-0.83) and 0.75(95%CI = 0.51-1.10), respectively, suggesting no appropriate genetic model was assessed using traditional method. Based on the model-free approach, significant result was found in the overall comparison  and among Asians, but not among Caucasians. Stratified by source of control, a significantly risk was found in the population-based studies, however, no significant association was observed in the hospital-based studies  and 0.46(95%CI = 0.11-2.00), respectively, suggesting a complete overdominant model was assessed using traditional method. Overall, no significant association with lung cancer risk was detected for MMP9 -1562 C/T polymorphism in complete overdominant model and heterogeneity between studies was observed in the overall comparison. In subgroup analysis based on ethnicity, however, a significantly decreased risk was found in Asians, suggesting homozygotes were at a lesser risk of lung cancer than heterozygotes. Based on the model-free approach, significant result was also found in Asians, suggesting lung cancer cases with higher mutational load than healthy individuals have higher risk for lung cancer susceptibility. Stratified by source of control, a significantly risk was found in the population-based studies, however, no significant association was observed in the hospital-based studies  and 1.32(95%CI = 1.03-1.67), respectively, suggesting a recessive model was assessed using traditional method. In recessive model, no significant association with lung cancer risk was detected for MMP13 -77A/G polymorphism in overall comparison and subgroup analysis  and MMP2 -1306 C/T in generalized odds ratio .Begg’s funnel plot and Egger’s test were performed to assess publication bias among the literatures. As shown in Meta-analysis is a powerful statistical tool to resolve the discrepancies across individual studies by integrating existing published data. At present, the majority of meta-analyses of genetic association studies are usually conducted by comparing genotype frequencies between cases and controls under various genetic models. However, these genetic models are not independent, and a priori knowledge or biological justification for model selection is seldom available3432In the current study, a total of 22 case-control studies with 8202 cases and 7578 controls were included in the meta-analysis1215161718192021222324252627282930in vitro with MMP13 -77A/G constructs indicated that the constructs with A had about twice as much transcriptional activity as the constructs with G in the same location1314151642This finding may be biologically plausible. MMPs play roles in many important physiological and pathological processes including cancer and lung inflammation through degradation of basal membranes and extracellular matrix242524252626272813143639et al conducted a comprehensive meta-analysis about five MMP polymorphisms and lung cancer susceptibility, and found that the MMP1-1607 1G/2G and MMP2-1306 C/T confer significantly susceptibility to lung cancer, and MMP1-1607’s effect was dependent on ethnicity, consistent with the results of this meta-analysis748212829et alet al et al MMP polymorphisms and lung cancer risk have been investigated by several meta-analyses444546Some heterogeneity factors between studies that could limit the strengths of the meta-analysis should be addressed. First, ethnicity was one of the most important factors that could lead to heterogeneity because of the diverse genetic backgrounds and environmental factors in different ethnicities. Second, the source of the controls was another factor that could lead to heterogeneity. Population-based controls could be more reliable than hospital-based controls because the genotype distributions in hospital-based controls may be deviated from normal. In this study, significant heterogeneity was found in three of the five polymorphisms. For these polymorphisms, the heterogeneity disappeared in subgroup analysis based on ethnicity, suggesting that ethnicity of the studied population are the major source of the heterogeneity.The current study has some inevitable limitations that should be acknowledged. First, only published studies were included in this meta-analysis, which may have biased our results. Second, there was significant heterogeneity among included studies. Even though we used the random-effect model to calculate pool ORs, the precision of outcome would be affected. Third, our results were based on an unadjusted estimated, a more precise analysis would have been conducted if more detailed individual data were available.In summary, we concluded that the MMP1-1607 1G/2G, MMP2-1306 C/T, MMP2-735 C/T, MMP9 -1562 C/T polymorphisms were risk factors for lung cancer among Asians, while MMP13 -77A/G polymorphism was not associated with lung cancer risk. However, future well designed large studies, particularly stratified by gene-gene and gene-environment interactions might be necessary to clarify the possible role of the MMP polymorphisms in the susceptibility to lung cancer.How to cite this article: Li, H. et al. Association of matrix metalloproteinase family gene polymorphisms with lung cancer risk: logistic regression and generalized odds of published data. Sci. Rep.5, 10056; doi: 10.1038/srep10056 (2015).


LIG4 gene may influence DNA repair ability, thus altering the genetic stability and resulting in carcinogenesis. A growing number of studies have investigated the relevance of LIG4 T9I (rs1805388) and D501D (rs1805386) polymorphisms with cancer risk, however, the results are conflicting. To obtain a comprehensive conclusion, we searched relevant literatures from PubMed, Web of Science, Ovid and Embase databases on May 15, 2014 and performed a meta-analysis. In this meta-analysis, a total of 17 articles were included. Of them, there were 15 studies with 5873 cases and 5771 controls for rs1805388 and 6 studies with 4161 cases and 4881 controls for rs1805386. Overall, our results suggested that there was no obvious relevance of LIG4 T9I polymorphism with cancer susceptibility. However, in subgroup analysis, we found the LIG4 T9I was associated with a slightly decreased cancer risk among Caucasians. As to the rs1805386, the genetic variant had no significant association with cancer risk. In conclusion, despite several limitations, this meta-analysis suggested that LIG4 T9I genetic variant is associated with a decreased risk of cancer among Caucasians, however, the rs1805386 gene polymorphism is not a risk factor of cancer.Polymorphisms of  Cancer is one of the most common causes of death in the world and results in a serious problem to global health1LIG4 gene, located on human chromosomes 13q33-34, encodes an ATP-dependent DNA ligase that joins single-strand breaks in a double-stranded polydeoxynucleotide in an ATP-dependent reaction6LIG4 is supposed to contribute to genomic instability and tumorigenesis.LIG4 variants and susceptibility of tumor. Then, we found that numerous studies investigated this issue, however, the conclusions are inconsistent. To obtain a comprehensive conclusion, we carried out a meta-analysis to systematically evaluate the relevance of LIG4 genetic variants with susceptibility of cancer.According to supposition mentioned above, we reviewed the related studies concerning of LIG4 polymorphism but not T9I or D501D252627282931323334353637383940414243444546A total of 167 papers were indentified from the databases as we described above. After deleting the duplications, 111 papers were left. Then we estimated the rest articles and 72 articles were discarded because of irrelevance with this issue. And one other potential eligible paper was obtained by screening the references of reviews. Of the remains 40 papers, one article was excluded for animal study9101112131517181920212223Of these included articles, one article353919Then, we established a database concerning of the information extracted from each included paper. Summaries of these studies were presented in LIG4 D501D variants with overall cancer risk  (We found that there was no obvious relevance of = 0.229) . In the LIG4 T9I variants with overall cancer risk . However, in the subgroup analysis, a statistically significant association was found among Caucasians .As shown in LIG4 D501D polymorphism and overall cancer susceptibility  and T9I . Thus, we selected the random-effect models to generate pooled ORs and corresponding 95% CIs for these models. On the other hand, no heterogeneity was observed among the other two models for D501D  and the fixed-effect models were performed to generate ORs and 95% CIs for them. The sensitivity analysis suggested that no obvious changes were observed for the pooled ORs when single investigation was excluded respectively (data were not shown).Heterogeneities were observed among several studies for P = 0.530; recessive: P = 0.919; dominant: P = 0.482; allele: P = 0.585) and D501D , which suggested that there were no significant publication bias among all these studies.The shape of the funnel plot showed that no evidence of asymmetry was observed in the current meta-analysis by Egger's test for T9I  were used to investigate the relevance of rs1805388 gene polymorphism with cancer susceptibility. Using the TSA (taking the data of dominant model for example), the required information size is 21516 subjects to demonstrate the issue . Until nGenomic instability is an outstanding characteristic of cancerIn eukaryotes, homologous recombination and NHEJ are two major pathways for DNA DSB repair and the latter is predominantly way in mammalian cellsLIG4 D501D polymorphism was associated with a decrease in breast cancer risk and Roddam et al.LIG4 T9I polymorphism may modulate predisposition to multiple myeloma. After these discoveries, a mass of investigations were performed to estimate the association between LIG4 T9I and D501D polymorphisms and the susceptibility of various cancers. To our puzzled, the results are conflicting. Thus, we performed a meta-analysis to obtain a comprehensive conclusion.Previous investigation had demonstrated that polymorphisms in DNA repair genes might alter DNA repair capacity and thus contribute to cancer riskLIG4 T9I and D501D polymorphisms and susceptibility of overall cancer. In this study, we found that rs1805386 and rs1805388 genetic variants had no relevance with overall cancer risk in homozygous, recessive, dominant and allele models. To further investigate the associations, we performed the subgroups analysis based on ethnicity and cancer types and found that rs1805388 variant is a decreased risk of cancers among Caucasians.To our knowledge, this is the first time to systematically estimate the associations between In order to make the conclusion more credible, we performed the publication bias analysis and sensitivity analysis according to Cochrane protocol. Funnel plots suggested that no obvious publication bias was observed. The sensitivity analysis indicated that the results are robust and no single study yield to obvious effects on the pooled ORs and the corresponding CIs. Besides, we performed the TSA and the results of TSA showed that the conclusions in this meta-analysis are robust.However, several limitations in this meta-analysis should be noticed. Firstly, several studies had small sample sizes which might lessen the statistical power. Secondly, the heterogeneity was existed and thus we performed the random-effects model to obtain the wider CIs, which might weaken the reliability of conclusions. Thirdly, our results were based on unadjusted assessment of ORs, which might influence the results. Fourthly, we did not search the unpublished studies, which might miss the relevant studies. Besides, all data included in this study were from published investigations which were based on the current marker identification method of the ‘one-step-clustering'. This approach might tend to be ‘passenger signals' instead of ‘drivers' and bury the ‘real' cancer gene, which made the results less robust and accurate51LIG4 D501D polymorphism has no obvious relevance with cancer risk and individual with LIG4 T9I genetic variant has a decrease cancer risk among Caucasians. With the development of research methods, future studies focusing on a combinatorial use of the polygene markers and integrative network modules analysis, are necessary to make the conclusions more comprehensive.Overall, in spite of these limitations, this analysis reached a precise conclusion that LIG4 or LIG 4 or ligase IV, 2) polymorphism or variant or variation or allele or genotype, and 3) cancer or carcinoma or tumor or neoplasm. And the last research was updated on May 15, 2014. All searched studies were screened and their references were retrieved to obtain other related articles. Then we downloaded the relevant papers and further screened to identify potentially eligible studies.We searched the PubMed, Web of Science, Ovid and Embase databases without language limitations for all related papers using the following searching strategies: 1) LIG4 polymorphisms (rs1805386 and rs1805388) and cancer susceptibility; (2) case-control design; (3) sufficient data provided to assess odds ratios (ORs) and the corresponding confidence intervals (CIs); (4) when multiple publication from a particular research group reported data from overlapping samples, the study reporting the largest or latest dataset was included. Exclusion criteria: (1) review articles; (2) case reports, or case-only studies; (3) studies that estimated the risk of prognosis.Studies included in this studies had to meet the following inclusion criteria: (1) estimating the relevance of All data were independently reviewed and extracted from the included papers by two authors (S.X. and J.H.). Disagreements were solved by full discussion until a consensus was reached. The following characteristics were collected from each study: first author's surname, year of publication, ethnicity, country, cancer type, sample size, control source, matching contents, the Hardy-Weinberg equilibrium, genotype methods and genetic distribution of cases and controls. The subgroup analysis was performed by cancer type and ethnicity.P value less than 0.05 levelP > 0.10 for the Q-test), the fixed-effects model (the Mantel-Haenszel method) was performed to calculate the combined OR2I statistics. The 2I value ranges from 0 to 100% and a larger 2I value indicating a greater heterogeneityAll the data management and analysis for this meta-analysis were performed with STATA 11.0 software . The strength of association between rs1805386 and rs1805388 polymorphisms and cancer susceptibility was estimated by calculating OR with the corresponding 95% CI. In order to calculate heterogeneity of studies, the Chi-Square test was used and significance was set at According to Cochrane Handbook for systematic reviews of interventions, meta-analyses and systematic reviews are considered to be the best available evidence if all available trials are included. However, ‘the best available evidence' might not be equal to ‘strong evidence' or ‘sufficient evidence'. It is well-known that meta-analysis may cause random errors in the series of sparse data and reduplicative testing on accumulating data. Based on these problems mentioned above, we applied the TSA to minimize the random errors and increase the robustness of conclusions58All authors contributed significantly to this work. Z.-G.C. designed and revised the article. J.H. collected articles, summarized data and revised this article. S.X. collected articles, summarized data, did statistical work and drafted the manuscript. X.-F.S. collected articles, summarized data and did statistical work. H.X. and K.S. provided collected data and summarized information in part. All authors reviewed this manuscript and approved the final draft.