In subgroup analyses, the positive association between total dairy product consumption and the risk of NHL was found among case-control studies  but not among cohort studies . The pooled RRs (95% CIs) of NHL were 1.21  for milk consumption in studies conducted in North America, and 1.24  for cheese consumption in studies that adopted validated food frequency questionnaires. In further analysis of NHL subtypes, we found statistically significant associations between the consumption of total dairy product  and milk  and the risk of diffuse large B-cell lymphoma. The dose-response analysis suggested that the risk of NHL increased by 5% (1.05 (1.00–1.10)) and 6% (1.06 (0.99–1.13)) for each 200 g/day increment of total dairy product and milk consumption, respectively. This meta-analysis suggested that dairy product consumption, but not yogurt, may increase the risk of NHL. More prospective cohort studies that investigate specific types of dairy product consumption are needed to confirm this conclusion.Many epidemiologic studies have explored the association between dairy product consumption and the risk of non-Hodgkin lymphoma (NHL), but the results remain controversial. A literature search was performed in PubMed, Web of Science and Embase for relevant articles published up to October 2015. Pooled relative risks (RRs) with 95% confidence intervals (CIs) were calculated with a random-effects model. The dose-response relationship was assessed by restricted cubic spline. A total of 16 articles were eligible for this meta-analysis. The pooled RRs (95% CIs) of NHL for the highest  Dairy product is an important part of diet in many countries around the world. Itcontains many essential nutrients, such as fats, proteins, minerals, vitamin D, and other bioactive nutrients. Dairy product can increase the content of total body and lumbar spine bone mineral in children  and decrNHL refers to a heterogeneous group of malignant tumors of lymphoid tissue that differs from Hodgkin disease. Over the past few decades, the incidence and mortality of NHL has been increasing internationally ,8,9. It We conducted a literature search to identify relevant available articles published in English from PubMed, Web of Science and Embase up to October 2015. Search terms included “dairy” (or “milk” or “butter” or “cheese” or “yogurt” or “ice cream”) and “non-Hodgkin lymphoma” (or “non-Hodgkin’s lymphoma” or “NHL”). We also reviewed the reference lists of the included studies for undetected relevant studies.The inclusion criteria are as follows: (1) case-control or cohort study published as an original study; (2) the exposure of interest were total dairy product, milk, butter, cheese, yogurt or ice cream; (3) the outcome of interest was non-Hodgkin lymphoma; (4) relative risk (RR) with 95% confidence interval (CI)  were provided; (5) the most recent and complete study was selected if data from the same population had been published more than once.Two investigators (Jia Wang and Xutong Li) searched and reviewed all identified studies independently. If the two investigators disagreed about the eligibility of an article, it was resolved by consensus with a third reviewer (Dongfeng Zhang).versus lowest category of the consumption of total dairy product and specific types of dairy product, and variables adjusted for in each studies.The following information was extracted from each study by two investigators independently: first author’s name, publication year, country in which the study was conducted, study design, follow-up duration, age range or mean age at baseline, sample size and number of cases, dietary assessment method, the type of dairy product, RR  with 95% CI for the highest i.e., g/1000 kcal), we estimated the absolute intakes by the mean energy intake of the participants [For dose-response analysis, the number of cases and participants (person-years), and RR (95% CI) for each category of total dairy product and milk were extracted. The median or mean level of total dairy product and milk for each category was assigned to the corresponding RR for every study. If the upper boundary of the highest category was not provided, we supposed that the boundary had the same amplitude as the contiguous category. If intakes were reported in densities  [P-values from meta-regression were calculated with a permutation test of 1000 to control the spurious findings [Pooled measure was calculated as the inverse variance-weighted mean of the logarithm of RR with 95% CI to assess the strength of associations between the consumption of total dairy product and specific types of dairy product and the risk of NHL, respectively. The DerSimonian and Laird random effect model (REM) was used to combine study-specific RRs (95% CIs) . The I2 ctively) . Meta-refindings . Subgroufindings . Small-sfindings .p-value for nonlinearity was calculated by testing the null hypothesis that the coefficient of the second spline is equal to 0.For dose-response analysis, a two-stage, random-effects, dose-response meta-analysis  was perfp-values) were two-sided with p < 0.05 considered statistically significant.All statistical analyses were performed with STATA version 12.0 . All reported probabilities  to assess the dietary consumption, and others used FFQs. With regard to study design, 13 articles were case-control studies, and three were cohort studies. The detailed characteristics of the included studies are shown in The main results are summarized in vs. lowest category of total dairy product consumption, the pooled RR of NHL was 1.20  and 1.02 , respectively .Seven articles ,21,22,28ectively . In subgpnonlinearity = 0.91), and the RRs (95% CIs) of NHL were 1.03 (0.98–1.08), 1.10 (0.95–1.27), 1.19 (1.00–1.41), 1.27 (1.06–1.51) and 1.42 (1.06–1.89) for 150, 390, 730, 1000 and 1500 g/day compared with 40 g/day, respectively. In addition, the dose-response analysis suggested that NHL risk increased by 5% (1.05 (1.00–1.10)) for each 200 g/day increment of total dairy product consumption  and 0.91 , respectively . In subgroup analysis stratified by dietary assessment method, the positive association was statistically significant in studies that adopted validated FFQs , but not in studies that used FFQs that had not been validated. In the further analysis of NHL subtypes, we just found a statistically significant association between milk consumption and the risk of DLBCL .Fourteen articles ,26,27,28ectively . In subgPnonlinearity = 0.78), and the RRs (95% CIs) of NHL were 1.04 (0.97–1.12), 1.07 (0.96–1.19), 1.11 (0.99–1.24), 1.12 (1.00–1.26) and 1.13 (1.00–1.28) for 120, 210, 370, 440 and 490 g/day compared with 0 g/day, respectively. In addition, the dose-response analysis suggested that NHL risk increased by 6% (1.06 (0.99–1.13)) for each 200 g/day increment of milk consumption  for the highest vs. lowest category of consumption. In subgroup analysis stratified by study design, no association was found in both case-control studies and cohort studies. In subgroup analysis stratified by continent in which the studies were conducted, the positive association was statistically significant only among studies conducted in Europe . In subgroup analysis stratified by dietary assessment method, cheese consumption was associated with an increased risk of NHL  in studies that adopted validated FFQs. In the further analysis of NHL subtypes, we did not find a statistically significant association between cheese consumption and any NHL subtypes.Nine articles ,25,26,27I2 = 36.9%, pheterogeneity = 0.190) for the highest vs. lowest category of consumption. For yogurt consumption, four articles [I2 = 81.6%, pheterogeneity = 0.001) for the highest vs. lowest category of consumption. In the further analysis of NHL subtypes, we did not find a statistically significant association between yogurt consumption and any NHL subtypes. For ice cream consumption, four articles [I2 = 72.3%, Pheterogeneity = 0.013) for the highest vs. lowest category of consumption.For butter consumption, four articles ,20,21,26articles ,20,21,24articles ,20,21,28p = 0.011). In the analysis of cheese consumption and the risk of NHL, the dietary assessment method was found to contribute to the between-study heterogeneity (p = 0.018). None of these covariates was found to have a significant impact on the between-study heterogeneity in other analyses.In order to explore the between-study heterogeneity, we performed univariate meta-regression with the covariates of sex, publication year, continent in which the studies were conducted, study design, dietary assessment method and whether the RR (95% CI) was adjusted for energy intake, smoking, alcohol intake and education. In the analysis of total dairy product consumption and the risk of NHL, study design was found to contribute to the between-study heterogeneity , milk (p = 0.616), butter (p = 0.798), cheese (p = 0.278), yogurt (p = 0.196) and ice cream (p = 0.250) and the risk of NHL. The funnel plot of the analysis of milk consumption and the risk of NHL was shown in the Egger’s test showed no evidence of a significant small-study effect for the analyses between the consumption of total dairy product ) and 6% (1.06 (0.99–1.13)) for each 200 g/day increment of total dairy product and milk consumption, respectively.3, the active form of vitamin D, had an anti-proliferative and pro-differentiation effect on cells that possess vitamin D receptors in follicular lymphoma cell lines [The mechanisms underlying the association between dairy product consumption and the risk of NHL have been postulated from several aspects. Dairy product is rich in protein and fat. Some experimental evidence from animal studies suggested that excessive intake of protein could give rise to chronic hyperstimulation of the immune system . Severalll lines . Howeverll lines . So far,ll lines ,53.Between-study heterogeneity is common in meta-analysis . It is nTo our knowledge, this is the first meta-analysis to explore the associations between the consumption of total dairy product and specific types of dairy product and the risk of NHL. A major strength of this meta-analysis is the large number of cases included, increasing the statistical power of the study to detect the associations. Second, RRs that reflected the greatest degree of control for potential confounders were extracted, indicating that the results were more credible. Third, considering the potential differences of components in different types of dairy product, we further assessed the effects of specific types of dairy product on the risk of NHL. Fourth, dose-response analysis was conducted to explore the relationships between the consumption of total dairy product and milk and the risk of NHL quantitatively. Fifth, almost all the included studies used a validated FFQ, which ensured the credibility of dietary assessment.However, there are some limitations in this meta-analysis. First, our results mainly came from case-control studies. There were only three cohort studies included in this meta-analysis, and the positive association between total dairy product consumption and the risk of NHL was not statistically significant in cohort studies. In addition, two other cohort studies ,56 withoIn summary, this meta-analysis suggested that dairy product, but not yogurt, may increase the risk of NHL. The risk of NHL increased by 5% and 6% for each 200 g/day increment of total dairy product and milk consumption, respectively. The results mainly came from case-control studies, and thus more cohort studies focusing on specific types of dairy product consumption are needed to confirm the conclusion.


CYP1A2 genes have the potential to affect the individual capacity to convert pre-carcinogens into carcinogens. With these comprehensive meta-analyses, we aimed to provide a quantitative assessment of the association between the published genetic association studies on CYP1A2 single nucleotide polymorphisms (SNPs) and the risk of cancer.Polymorphisms in the CYP1A2 SNPs and cancer. Random effect model was used to calculate the pooled ORs. Begg and Egger tests, one-way sensitivity analysis were performed, when appropriate. We conducted stratified analyses by study design, sample size, ethnicity and tumour site.We searched MEDLINE, ISI Web of Science and SCOPUS bibliographic online databases and databases of genome-wide association studies (GWAS). After data extraction, we calculated Odds Ratios (ORs) and 95 % confidence intervals (CIs) for the association between the retrieved P for heterogeneity < 0.01; I2 = 50.4 %). When stratifying for tumour site, an OR of 0.84  was reported for bladder cancer for those homozygous mutant of rs762551. An OR of 0.79  was obtained for the bladder cancer from the hospital-based studies and on Caucasians.Seventy case-control studies and one GWA study detailing on six different SNPs were included. Among the 71 included studies, 42 were population-based case-control studies, 28 hospital-based case-control studies and one genome-wide association study, including total of 47,413 cancer cases and 58,546 controls. The meta-analysis of 62 studies on rs762551, reported an OR of 1.03  for overall cancer  for bladder cancer for those homozygous mutant of rs762551. Due to the limitations of our meta-analyses, the results should be interpreted with attention and need to be further confirmed by high-quality studies, for all the potential CYP1A2 SNPs.This large meta-analysis suggests no significant effect of the investigated  Cancer is a complex disease that develops as a result of the interactions between environmental factors and genetic inheritance. In 2012 there were 14.1 million new cancer cases and 8.2 million cancer deaths worldwide .CYP1A2 gene  AND (Cancer) AND (Humans [MeSH]), without any restriction on language. All eligible studies were retrieved, and their bibliographies were hand-searched to find additional eligible studies. We only included published studies with full-text articles available.Identification of the studies was carried out through a search of MEDLINE, ISI Web of Science and SCOPUS databases up to February 15th, 2015 for the association between CYP1A2 and various cancers using the combinations of following terms: (Cytochrome P450 1A2) OR (CYP1A2) OR (Chromosome 15q24.1) AND (Cancer). Additional consultation of principal investigators (PI) of the retrieved GWAS was undertaken in order to obtain the primary data and include them in the analyses.Also, detail search of several publically available databases of genome-wide association studies (GWAS) - GWAS Central, Genetic Associations and Mechanisms in Oncology (GAME-ON), the Human Genome Epidemiology (HuGE) Navigator, National Human Genome Research Institute , The database of Genotypes and Phenotypes (dbGaP), The GWASdb, VarySysDB Disease Edition (VaDE), The genome wide association database (GWAS DB), was carried out up to February 15CYP1A2 gene polymorphism in relation to the number of cancer cases and controls, according to the three variant genotypes (wild-type homozygous (wtwt), heterozygous (wtmt) and homozygous mutant (mtmt)). Case-only and case series studies with no control population were excluded, as well as studies based only on phenotypic tests, reviews, meta-analysis and studies focused entirely on individuals younger than 16 years old. When the same sample was used in several publications, we only considered the most recent or complete study to be used in our meta-analyses. Meanwhile, for studies that investigated more types of cancer, we counted them as individual data only in a subgroup analysis by the tumour type, while when they reported different ethnicity or location within the same study, we considered them as a separate studies.Studies were considered eligible if they were assessing the frequency of any CYP1A2 polymorphisms in the compared groups. We used widely accepted National Center for Biotechnology Information (NCBI) CYP classification [Two investigators (C.I. and V.V.) independently extracted the data from each article using a structured sheet and entered them into the database. The following items were considered: rs number, first author, year and location of the study, tumour site, ethnicity, study design, number of cases and controls, number of heterozygous and homozygous individuals for the fication  to deterCYP1A2 SNP and cancer were defined as follows:1)wtmt vs wtwt (OR2).mtmt vs wtwt (ORThe estimated Odds Ratios (ORs) and 95 % confidence interval (CI) for the association between each 2 ≠ 1 and OR1 = 1Recessive model (mtmt versus wt carriers): if OR2 = OR1 ≠ 1,Dominant model (mt carriers versus wtwt): if ORAccording to the following algorithm on the criteria to identify the best genetic model  for each2-based Q statistics and the I2 statistics [I2 = 0 % indicates no observed heterogeneity, within 25 % regarded as low, 50 % as moderate, and 75 % as high [http://ihg.gsf.de/cgi-bin/hw/hwa1.pl ). Additionally, the Galbraith’s test [we used the dominant model of inheritance for rs2069514, rs2069526 and rs35694136 and recessive model for rs762551, rs2470890 and rs2472304 in the meta-analysis. Random effect model was used to calculate the pooled ORs, taking into account the possibility of between studies heterogeneity [ as high . A visua as high  were use as high . To deteh’s test  was perfp-value ≤ 0.01 were excluded from the meta-analyses, given that this deviation may represent bias. We conducted stratified analysis by study design, ethnicity, sample size and tumour site to investigate the potential sources of heterogeneity across the studies. Statistical analyses were performed using the STATA software package v. 13 , and all statistical tests were two-sided.Studies whose allele frequency in the control population deviated significantly from the Hardy-Weinberg Equilibrium (HWE) at the We identified a total of 2541 studies through MEDLINE, ISI Web of Science and SCOPUS online databases. One thousand and sixteen studies were left after duplicates removal, and after carefully reading the titles, only 175 studies were assessed for eligibility. After reviewing the abstracts, 120 full text articles were obtained for further eligibility. By not fulfilling the inclusion criteria, 61 full text articles were excluded, leaving 59 studies for quantitative synthesis. Additional hand-search of the reference lists of 59 included studies was done and 11 new eligible studies were found, resulting in 70 included studies.CYP1A2 SNPs and cancer risk were identified after detail search of GWAS online databases. Studies did not report full data on investigated SNPs, so we contacted principal investigators (PIs) to retrieve the information and include into our analyses. After 3 repeated solicitations, only one PI provided us with the full data on CYP1A2 SNPs of breast cancer cases and controls, and by this making total of 71 studies included in our meta-analyses [Eleven GWASs on the association between analyses , 61–84. Among the 71 included studies, 42 were population-based case-control studies, 28 hospital-based case-control studies and one genome-wide association study, including total of 47,413 cancer cases and 58,546 controls  and an OR2 of 1.06 (95 % CI 0.97–1.16), for rs2470890 OR1 1.03 (95 % CI 0.93–1.14) and OR2 of 1.14 (95 % CI 0.97–1.34) and for rs2472304 OR1 of 0.98 (95 % CI 0.79–1.22) and OR2 of 0.89 (95 % CI 0.66–1.22) according to the criteria proposed in the methods section, we applied the recessive model of inheritance for the meta-analyses. On the other hand, for rs2069514, rs2069526 and rs35694136 original papers did not report enough data to calculate OR1 and OR2, so we were able only to apply the dominant model for the data analyses.As the crude analysis for rs762551 provided an ORCYP1A2 SNPs and cancer. By pooling 62 studies on rs762551, the meta-analysis reported an OR of 1.03 (95 % CI 0.96–1.12) for overall cancer . Egger test and the Begg’s correlation method did not provide statistical evidence of publication bias  .The Figs. CYP1A2 on cancer risk for Caucasians , Asians  nor among a mixed population . When stratifying according to the tumour site, results showed an OR of 0.84  for bladder cancer for those homozygous mutant types of rs762551  for bladder cancer among the hospital-based population and among Caucasians. There was no significant association among Caucasians for breast cancer , lung cancer  or colorectal cancer . Among Asians, when stratifying for cancer type, we obtained an OR of 0.76  for colorectal cancer and OR = 1.27  for breast cancer.Results of the stratified meta-analyses are reported in the Table P for heterogeneity <0.01; I2 = 60 %) . We performed the Galbraith’s test to explore the source of heterogeneity and accordingly singled out the study of B’chir F. et al. [P = 0.28), with the OR of 0.93 (95 % CI 0.82–1.06).When pooling the 20 studies on rs2069514, the meta-analysis provided an OR of 0.99 (95 % CI 0.81–1.21) for overall cancer  for colorectal cancer, an OR of 1.29  for lung cancer . We did not observe any significant association between rs2069514 polymorphism and cancer risk when subgrouping data according to ethnicity, source of controls and sample size and then stratified by cancer type. Among Caucasians, we obtained an OR of 1.28  for lung cancer, while among Asians OR = 0.94  for lung and OR = 0.94  for colorectal cancer.We evaluated the effect of the rs2069514 polymorphism according to the tumour site and obtained an OR of 0.96  . The Galbraith’s test singled out the study of Anderson LN et al. [P = 0.39), with still not significant OR of 1.06 . The effect of rs2470890 polymorphism according to the tumour site was also evaluated and was obtained non-significant result of OR of 1.10  P for heterogeneity = 0.51, I2 = 0.0 % for colorectal cancer and an OR of 1.20 , P for heterogeneity = 0.09; I2 = 65.7 % for cancer of upper aero-digestive tract (UADT)  nor Asians .We performed meta-analysis of 11 studies on rs2470890 which provided an OR of 1.11 (95 % CI 0.96-1.28) for the overall cancer risk .Results of the remaining three SNPs of CYP1A2 SNPs and cancer risk  and heterocyclic aromatic amines (HAA) [CYP1A2*1 F represents a highly inducible genotype, and this high CYP1A2 activity may increase the hydroxylated forms as proximate carcinogens, from HCAs and aryl-amines [The es (HAA) . Therefoes (HAA) . The will-amines .CYP1A2 polymorphisms were significantly associated with overall risk of cancer at various sites. These results confirm the findings of a recent meta-analysis from Li Zhenzhen et al. [CYP1A2 rs762551 polymorphisms and the risk of cancer, to date.In our meta-analyses, we showed that none of the investigated n et al.  where waWhen stratifying according to tumour site, our results showed a borderline not significant OR of 0.84  for bladder cancer for those homozygous mutant of rs762551 with total of 3430 cases and 3242 controls included (Table We also found a significant decreased risk for bladder cancer for mutant carriers of rs762551 among the hospital-based population. Hospital-based studies have certain biases since those controls may have some benign diseases which can progress and also may not be representative of the general population. Using a population-based control would reduce the chance of bias in these studies.CYP1A2 rs762551, rs2069514, rs2069526, and rs2470890 polymorphisms and lung cancer risk, there was no evidence of significant association between lung cancer risk and CYP1A2 rs2069514, s2470890, and rs2069526 polymorphisms. They found increased lung cancer risk for rs762551 polymorphism in Caucasians from 3 studies, while in our analysis there was no such connection on a bigger sample of studies [In one recent meta-analysis by Zhi-Bin Bu et al.  on the a studies , 30–32.CYP1A2 rs762551 polymorphism was not a risk factor for colorectal cancer susceptibility, since no association was detected after all studies were pooled together nor in a subgroup analysis by ethnicity or source of controls, in all genetic models. The influence of the different CYP1A2 SNPs might be camouflaged by the presence of some yet unidentified causal genes involved in many other types of cancer.Lastly, when stratifying our results for breast and colorectal cancer, we did not report any significant association between rs762551 and these cancers, thus confirming previous meta-analyses of Li-Xin Qiu et al.  on breasWhen stratifying the results according to ethnicity, the protective effect of rs2472304 in our study was restricted only to Caucasians, while for rs2470890, we noticed an increased risk among a mixed population. A possible explanation for these results could be that the same polymorphisms may play different roles in cancer susceptibility in different ethnic populations as well as different tumour positions, due to a difference in genetic backgrounds, the environment they live in, lifestyle and migrations, which all may have a critical role in cancer pathogenesis . Also, sNo other relevant results were reported for the remaining SNPs, however there were available only few studies regarding these associations, involving relatively small number of participants.CYP1A2 polymorphisms was limited to perform any meaningful subgroup analyses. Thirdly, it would have been valuable to stratify the results according to environmental effect modifiers, though this was not possible, as the original data sets were not available. Indeed, due to lack of access to original data used in included studies, our meta-analyses are based on the unadjusted data, so the effects might be confounded or modified by relevant covariates. Fourthly, beside breast cancer, there are no genome-wide association studies of the effects of CYP1A2 polymorphisms on cancer risk. We were able to include only one breast cancer GWAS into our analyses, therefore our results might be affected by additional publication bias.In interpreting the results, some limitations of our study should be considered. Firstly, only published studies were included, so there was space for publication bias, which in fact was confirmed by formal statistical tests. Secondly, the study size for most of the CYP1A2 polymorphism and cancer risks and investigated 6 different CYP1A2 SNPs.Despite these limitations, our meta-analyses also have some advantages. First, the statistical power of the analyses was noticeably increased as a huge number of cases and controls were pooled from different studies and has more statistical powerful than any single case-control study. Secondly, in our analyses, we included more studies than any previously published meta-analysis on the association between CYP1A2 polymorphisms are not associated with cancer susceptibility under various genetic models. In order to reach a more definitive conclusion, there is a necessity for further gene-gene and gene-environment interaction studies to be conducted on different populations and larger sample size, for diverse CYP1A2 SNPs.In conclusion, our meta-analysis suggests that investigated


MMP3 -1171 5A/6A polymorphism (rs35068180) and periodontitis has been widely studied. However, existing studies have yielded contradictory results. We therefore conducted a meta-analysis to comprehensively investigate these inconclusive findings. Several electronic databases were searched for eligible articles. Seven case-control studies from 6 articles were searched without any language restrictions. Pooled estimates indicated that MMP3 -1171 5A/6A polymorphism is associated with a decreased risk of periodontitis . Similar results were also found in chronic periodontitis (CP), Asian, Asian&CP, and non-smokers subgroups. Moreover, MMP3 rs35068180 polymorphism might be associated with a lower risk of aggressive periodontitis (AgP) in Asians , and CP in Caucasians and Brazilians. In conclusion, this meta-analysis demonstrates that MMP3 -1171 5A/6A polymorphism may be associated with decreased risk of both CP and AgP in Asians. Large independent studies to replicate these results are necessary to validate these associations in other populations.Matrix metalloproteinase-3 (MMP3) plays a key role in tissue degradation in periodontitis. The relationship between the  IL-1, IL-8 and TGF-β1)78Periodontitis is a globally prevalent inflammatory disease that affects the tooth-supporting apparatus5MMP3 gene, located on chromosome 11q22.2–22.3 (adjacent to the MMP1 and MMP12 genes), is an endopeptidase produced by smooth muscle cells, macrophages and synovial cellsMMP3 promoter region (rs35068180 or rs3025058) can alter its transcription levelsMMP3 as one of the most promising genes to be involved in periodontitisMatrix metalloproteinases (MMPs) are the most important pathway in periodontitis-associated tissue breakdown, owing to their role in the pathological destruction of extracellular matrix and immune responses related to periodontal inflammationMMP3 -1171 5A/6A polymorphism and the risk of periodontitis. However, the results of these studies were inconsistent. The discrepancies among studies may be due to the relatively small sample size in each investigation, as well as varying population characteristics. Therefore, we performed a meta-analysis to investigate the potential association of MMP3 -1171 5A/6A polymorphism with the risk of periodontitis.Over the past decade, many genetic association studies have been conducted to evaluate the association between Seven case-control studies from six publications were identified in the meta-analysis , the other four genetic models contributed to decreased susceptibility to periodontitis . After the HWE-violating studies were excluded, a significant association was found in the allele contrast .Between-study heterogeneity was not evident in all five comparisons . OverallMMP3 promoter region with both CP  and AgP   .vs. 5A allele: OR = 0.68, 95% CI = 0.58–0.80, Pheterogeneity = 0.149; 5A/6A vs. 5A/5A: OR = 0.52, 95% CI = 0.40–0.68, Pheterogeneity = 0.198; 6A/6A vs. 5A/5A: OR = 0.45, 95% CI = 0.25–0.81, Pheterogeneity = 0.173; 5A/6A + 6A/6A vs. 5A/5A: OR = 0.51, 95% CI = 0.39–0.66, Pheterogeneity = 0.202, respectively), Caucasians  , and Brazilian mixed populations    . Furtherctively) .vs. 5A allele: OR = 0.73, 95% CI = 0.60–0.88, Pheterogeneity = 0.414; 5A/6A vs. 5A/5A: OR = 0.54, 95% CI = 0.39–0.75, Pheterogeneity = 0.151; 6A/6A vs. 5A/5A: OR = 0.50, 95% CI = 0.28–0.89, Pheterogeneity = 0.260; 5A/6A + 6A/6A vs. 5A/5A: OR = 0.53, 95% CI = 0.39–0.73, Pheterogeneity = 0.189, respectively)  .Sensitivity analysis was performed by omitting individual studies to assess the effect of each publication on the overall results. No single study changed the pooled ORs qualitatively, which suggests that the results of our meta-analysis are accurate.Begger’s funnel plot and Egger’s test were performed to detect the publication bias of included studies. As shown in MMP1 -1607 1G/2G and MMP9 -1562C/T polymorphisms may be involved in the development of CP20MMP3 -1171 5A/6A and MMP1 -1607 1G/2G (rs1799750) are in a strong linkage disequilibriumMMP3 -1171 5A/6A polymorphism and periodontitis has been widely investigated. However, previous studies have yielded contradictory results. We therefore performed this meta-analysis to comprehensively evaluate these inconclusive findings. To the best of our knowledge, this is the first meta-analysis to have investigated the association between MMP3 -1171 5A/6A polymorphism and the susceptibility to periodontitis. Our meta-analysis was based on seven case–control studies that focused on the link between MMP3 -1171 5A/6A polymorphism and periodontitis risk and included a total of 1,213 cases and 1,831 controls. The results of our meta-analysis revealed that -1171 5A/6A polymorphism is associated with a decreased risk of overall periodontitis; this association was consistently significant in all subgroups in analyses that were stratified by periodontitis type, ethnicity and smoking status.Increasing evidence suggests that MMPs play a pivotal role in periodontitisTNF-α -308G/A and Fcγ receptor IIIB NA1/NA2) have been shown to have similar effects on AgP and CP24MMP3 -1171 5A/6A polymorphism on two forms of periodontitis. We found that this locus was associated with a decreased risk of periodontitis under four genetic models. Additionally, -1171 5A/6A showed a trend towards protective factor for periodontitis in the recessive model, although the finding was not statistically significant (p = 0.06). When the analysis was limited to the four HWE studies, the association between -1171 5A/6A polymorphism and periodontitis was still significant in the allele model, indicating the robustness of this meta-analysis. Several subgroup analyses were subsequently conducted, and the associations were similar in AgP and CP subgroups in the allele comparison. Additional well-designed studies are required to strengthen our understanding of the association between -1171 5A/6A polymorphism and AgP. Although the minor allele frequency (MAF) and genotype distributions in the included studies for -1171 5A/6A varied by ethnicity  increased in periodontitis cases and played a role in tissue destruction27MMP3 -1171 5A/6A polymorphism and systemic diseases29303132343536MMP3 rs35068180 variant. We speculate that rs35068180 may influence the development of periodontitis through a similar molecular mechanism. In vitro assays of promoter activity have shown that the 5A allele has a two-fold higher promoter activity than the 6A alleleThe role of The following limitations of the meta-analysis should be recognised. First, we pooled the data based on unadjusted information, while a more precise analysis needs to be conducted if individual data are available. Second, the number of subjects included in this meta-analysis was relatively small for AgP, Caucasian and Brazilian mixed population subgroups. Hence, the results should be interpreted with caution. Third, the genotypic distribution of the controls also deviated from HWE in three of the studies.MMP3 -1171 5A/6A polymorphism contributes to a decreased susceptibility to CP, especially in Asian populations. Furthermore, this polymorphism may decrease the risk of AgP in Asians. Replication studies of our results in large independent populations are necessary to validate this association.In conclusion, this meta-analysis suggests that MMP3 -1171 5A/6A polymorphism and periodontitis was performed. The literature was retrieved in electronic bio-medical databases including Google Scholar, PubMed, Embase, Medline and China National Knowledge Infrastructure (the last search update was 19 Apr 2015). A combination of the terms “Matrix Metalloproteinases”, “MMP”, “periodontal diseases”, “periodontitis”, and “polymorphism” were entered both as text words and as Medical Subject Heading components. No language restriction was placed on the search. Manual searches for references cited in published review and original publications were also performed.Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) guidelines were followedThe following two inclusion criteria were used for all retrieved articles: (1) use of a case–control, cohort, nested case–control, or cross-sectional design; and (2) genotype distribution or allele frequency was available from the published data for both cases and controls. The exclusion criteria were as follows: (1) lack of usable allele frequency or genotype distribution data; and (2) lack of control subjects.Information was carefully extracted independently from all eligible publications by two investigators. Disagreements were resolved through discussion among the authors to achieve a consensus. The following characteristics were collected for each study: the first author’s family name, publication year, ethnicity , numbers of cases and controls, type of periodontitis (CP or AgP), distribution of genotypes and alleles, smoking status, and genotype identification method.The quality of each study was evaluated independently by two authors according to the modified STROBE quality score systemsMMP3 gene -1171 5A/6A polymorphism and periodontitis risk. The Z-test was used to determine the statistical significance of the pooled OR. The pooled ORs were performed for the following genetic models: the mutant allele versus the wild type, homozygote contrast, heterozygote contrast, dominant model and recessive model. In addition, the HWE in the healthy control populations was tested within each study. We applied the chi-squared method to assess if the genotype distribution in the control group was in HWE. Subgroup analyses were also performed for the HWE in controls, type of periodontitis, ethnicity and smoking status.Odds ratios (ORs), along with 95% confidence intervals (CIs), were used to assess the strength of the association between the I2 statistic. The fixed-effects model was used when heterogeneity was not significant (Pheterogeneity > 0.10 or I2 < 50%); otherwise, the random-effects model was applied. Sensitivity analysis was performed to assess the stability of the association. Begg’s funnel plotp-values.Both the fixed-effects model (the Mantel–Haenszel method)How to cite this article: Ding, C. et al. Matrix Metalloproteinase-3 -1171 5A/6A Polymorphism (rs35068180) is Associated with Risk of Periodontitis. Sci. Rep.5, 11667; doi: 10.1038/srep11667 (2015).


MMP-9 promoter −1562 C/T polymorphism with the susceptibility and/or severity of chronic periodontitis (CP), results have been inconsistent. Therefore, a meta-analysis of all eligible studies was performed to derive a more precise estimation of the association between this polymorphism and CP risk.Although many studies have focused on the association of the MMP-9 promoter −1562C/T polymorphism with CP risk using validated genotyping methods, and with non-duplicated data were selected for this study. A fixed-effect model was used to calculate pooled ORs in the absence of heterogeneity across included trials (P > 0.1 and I2 < 50 %), otherwise the random-effect model was applied.All relevant studies were identified through a database search in PubMed, Medline, and Web of Science. All the full-text studies with appropriate analytical design, published in English, which evaluated the association of MMP-9 promoter −1562 C/T polymorphism was associated with a significantly decreased risk for CP under all comparison models. In subgroup analyses by ethnicity, pooled ORs showed that a significant association of the MMP-9 promoter −1562 C/T polymorphism with CP risk was only detected in Caucasians and Asians but not in mixed population. In the subgroup analysis by severity of CP, pooled ORs indicated that T allele of the MMP-9 promoter −1562 C/T polymorphism was associated with decreased susceptibility to severe CP while there was no significant association between this polymorphism and moderate CP.In an overall meta-analysis, pooled ORs revealed that T variant in the MMP-9 promoter −1562 C/T polymorphism might be a protective factor for CP, especially in Caucasians and Asians. Moreover, there was a significant association of the MMP-9 promoter −1562 C/T polymorphism with decreased susceptibility to severe CP, while the allelic and/or genotype distributions of this polymorphism were not associated with moderate CP.Our meta-analysis showed that T allele in the The online version of this article (doi:10.1186/s40064-016-2135-3) contains supplementary material, which is available to authorized users. Chronic periodontitis (CP), the most commonly occurring and slowly progressive form of periodontal disease, can lead to continual inflammatory host response, which may finally result in periodontal attachment loss and bone resorption  belong to a family of zinc-dependent endopeptidases, which are served as the vital enzymes. They have the combined capacity to degrade extracellular matrix (ECM) components such as several kinds of collagen, that are essential for the tissue repair and remodeling involved with the development of inflammation  or gelatinase B, involved in the breakdown of various connective tissue proteins, including types IV, V and XI collagen, proteoglycans as well as elastin, is abundantly expressed in CP  exists in MMP-9 gene at position −1562, which affects transcription, and also decreased transcriptional activity shown by CC genotype is well documented  of PubMed, Medline, and Web of Science, with combinations of the following terms used as MESH headings and free text words:  and (“genetic variant” or “genetic variation” or “polymorphism”) and . We limited all searches to clinical trial, meta-analysis as well as review. In addition, any potentially relevant papers that may have been missed during the process of computer-assisted searches were also identified via the manual search of bibliography lists.MMP-9 promoter −1562 C/T polymorphism with CP risk among CP affected and unaffected individuals; (2) Studies applied validated genotyping methods such as PCR–RFLP; (3) Studies with appropriate analytical design, for example case–control, cohort, or nested case control; (4) Studies published in English, and available full-text; (5) Study data not duplicated or overlapped with those of any other article. Besides, we barred those studies, of which the pertinent data were not available to figure up the odds ratios (ORs) and its variance.The following criteria were designed and used for including the identified studies into the present meta-analysis and system review: (1) Studies that evaluated the association of http://ihg2.helmholtz-muenchen.de/cgi-bin/hw/hwa1.pl).The data extraction was performed by two independent reviewers (ADH and PS) under a pre-defined strategy. Any disagreements between investigators were settled through consensus decision with the third evaluator (LWY). The following items were collected from each included trial: first author’s surname, publication year, nation, race, sample size, severity of CP, matching criteria, genotyping method, as well as the results of HWE in controls were calculated using an online software . The Z-test was applied to find out statistical significance of pooled ORs. Initially, the allele-frequency comparison model (T vs. C) was used to evaluate the potential relationship between the allele T of MMP-9 promoter −1562 C/T polymorphism and CP risk. After that, the association of this polymorphism with CP risk was also examined by four other comparison models, including TT versus CC, CT versus CC, dominant (TT + CT vs. CC), as well as recessive (TT vs. CT + CC) genetic models. In addition, subgroup analyses concerning the study characteristics of ethnicity and severity of CP were respectively carried out.The association of the I2 < 50 %), otherwise, the random-effects model was applied. Furthermore, the potential source of heterogeneity was evaluated via a sensitivity analysis. Besides, we used funnel plots to examine publication bias, in which a standard error of log (OR) of each included trial was plotted against its log (OR), and Egger’s linear regression test was applied to assess the asymmetry. Meta-analyses was executed by using software Stata , with a statistical significance defined as two-sided P value <0.05. If the results of trials could not be pooled through the meta-analysis, we assessed them using descriptive statistics.To calculate pooled ORs, a fixed-effects model was planned to be used with the absence of heterogeneity across included trials , but with a considerable heterogeneity among the included studies    Table . Besides0) Table , pooled 0) Table .Table 3SMMP-9 promoter −1562 C/T polymorphism  occurs  in the promoter region of Moreover, it has been proved that inappropriate activation of MMP-9 at a site of infection can have a profound effect in the tissue destruction of CP , Latin-European whites, mulattoes (black-white mix), zambos (black-Amerindian mix), blacks (Afro-Americans), and pure indigenous Amerindians (de Souza et al. In the subgroup analysis by ethnicity, pooled ORs revealed that a significant association of the MMP-9 promoter −1562 C/T polymorphism was associated with decreased susceptibility to severe CP while there was no significant association between this polymorphism and moderate CP. Our results were in accordance with the previous studies (Gurkan et al. MMP-9 promoter −1562 C/T polymorphism not only contributes to the disease susceptibility but also the severity of CP.In the subgroup analysis by severity of CP, the results indicated that T allele of the Most of the evidence from our study should be considered to be stable and convincing albeit substantial heterogeneity since the heterogeneity detected in many pooled analyses did not have a significant impact on the results of our study. However, some potential limitations still exist in our study that needs to be addressed before interpreting these results. Firstly, a relatively large heterogeneity was evident in this meta-analysis. Nevertheless, through stratified analysis by ethnicity and severity of CP, heterogeneity reduced significantly. As a result, we presumed that relatively large heterogeneity mainly results from differences in ethnicity and severity of CP. Secondly, our overall outcomes were based on individual unadjusted ORs. A more precise analysis would have been conducted by adjustment of other covariates, such as age, gender, smoking status, environmental factors, and etc. thereby avoiding serious confounding bias. Accordingly, control group individuals might develop CP later within the age range, if unmatched by age and gender, as seen in some studies(de Souza et al. MMP-9 promoter −1562 C/T polymorphism might be a protective factor for CP, especially in Caucasian and Asian populations. Moreover, there was a significant association of the MMP-9 promoter −1562 C/T polymorphism with decreased susceptibility to severe CP, while there was no association seen between the allelic and/or genotype distributions of this polymorphism and moderate CP. However, more studies of high quality with larger sample sizes and also other ethnic populations are still needed to draw a more solid inference on the relation between this polymorphism and CP risk.In conclusion, our meta-analysis showed that T allele in the


Pnonlinearity < 0.001). The SRR were 0.98 (95%CI: 0.95–1.01) for 25 kg/m2, 0.91 (95%CI: 0.85–0.98) for 30 kg/m2 and 0.81 (95% CI: 0.72–0.91) for 35 kg/m2, with mild between-study heterogeneity (I2 = 5%). The results of the stratified analysis by gender were comparable to those of the overall meta-analysis. When stratified by smoking status, linear dose-response associations were observed for current smokers, ex-smokers and non-smokers (Pnonlinearity > 0.05), whereas the effects were attenuated when restricting analysis to non-smokers, and at the point of 30 kg/m2, the SRR was 0.96 (95%CI: 0.86–1.07) for males and 0.95 (95%CI: 0.89–1.02) for females. This meta-analysis provides quantitative evidence that increasing BMI is a protective factor against LC. Keeping normal-to-moderate BMI should be prescribed as an evidence-based lifestyle tip for LC prevention in smokers.Questions remain about the significance of the dose-response relationship between body mass index (BMI) and lung cancer (LC) risk. Pertinent studies were identified through a search in EMBASE and PUBMED from July 2014 until March 2015. The summary relative risk (SRR) and confidence interval (CI) were estimated. The dose-response relationship was assessed using a restricted cubic spline. The overall meta-analysis showed evidence of a nonlinear association between BMI and LC risk ( Lung cancer (LC) is one of the most prevalent and deadliest human cancers. The average five-year survival rate after diagnosis is 16.3%2345789Several observational epidemiological studies have shown that higher BMI correlates with a lower risk of LC6781211917No work to date has clarified the dose-response relationship between BMI and LC risk, not to mention the differences relative to the classification of smoking status. In addition, the previous meta-analysis was performed simply by combining the same categories of BMI across individual studies and the resulting conclusions have been questioned. We therefore carried out a dose-response meta-analysis of prospective cohort studies to provide a more precise evaluation for the potential association between BMI and LC risk. We particularly wanted to clarify potential modification by smoking status, as well as to investigate potential heterogeneity by subgroup and meta-regression analyses.A comprehensive literature retrieval was conducted by using the databases PubMed (Medline) and Embase from 6th July 2014 until 13th July 2014. An updated secondary search was performed until 1st March 2015. Pertinent articles were identified by using the key words “body mass index,” “BMI,” “obesity,” “overweight,” “body size” or “body weight” combined with “lung cancer,” “lung neoplasm” or “lung carcinoma”. Additionally, we scrutinized references from relevant original papers and review articles to obtain other pertinent publications. No language restrictions were imposed. We followed the criteria and standard for conducting meta-analyses of observational studies and reporting the results2; (c) with cohort design; (d) investigators provided the following data: the number of cases, the total subjects or person-years and estimates of relative risk (RR)  with 95% confidence intervals (95%CIs) for three or more quantitative categories of BMI. If the person-years were not reported, they were calculated through multiplying the number of subjects by mean follow-up durationThree authors independently retrieved and assessed potentially relevant articles according to the prespecified selection criteria. Discrepancies between reviewers were solved by discussion. To be included, the study had to meet the following criteria: (a) published as an original article; (b) has definite description of BMI in kg/mTwo reviewers designed and pilot-tested an ad hoc developed excel sheet for data extraction, which was eventually approved by the authors’ team. General information, study characteristics, participants’ characteristics, results as reported , associated raw data for re-calculation (data checking) and any multivariate analyses adjustment factors (if applicable) were collected.For each study, the median or mean value for each category of BMI was assigned to each corresponding estimate of RR. When the median or mean value per category was not obtained in the study, we assigned the mid-point of the upper and lower boundaries in each category as the average BMI. For open-ended categories, value equal to 0.83 (1/1.2) times the upper limit of the lowest category and equal to 1.2 times the lower limit of the highest category were consistently usedTo assess study quality of the included studies, a 9-point scoring system according to the Newcastle-Ottawa Scale was usedIn this meta-analysis, the relative risks (RRs) and 95%CIs were considered as the effect size for all studies. Because the incidence of LC is low, the HR from cohort studies is approximated to RR. Besides, we just included RR or HR based on the Cox proportional hazard model for statistical analysis. Thus we reported all risk estimate produced by current meta-analysis as RR for simplicity. We extracted the RRs from the maximally adjusted model to lower the risk of possible confounding.2), normal (BMI = 18.5–24.9 kg/m2), overweight (BMI = 25–29.9 kg/m2), and obesity (BMI ≥ 30 kg/m2). Normal group (BMI = 18.5–24.9 kg/m2) was considered as reference group2 statistic We followed the WHO international classification and defined body mass categories as follows: underweight (BMI < 18.5 kg/m2 increase from BMI = 20 to 35 kg/m2) using the restricted maximum likelihood method in a multivariate random-effects meta-analysisP value for non-linearity was computed by testing the null hypothesis that the coefficient of the second spline was equal to zeroA dose-response meta-analysis was performed to examine a potential nonlinear relationship between BMI and LC. We used restricted cubic splines with 3 knots at fixed 10%, 50% and 90% to model a potential curvilinear relationship and then a dose-response meta-analysis was conducted using the generalized least-squares trend command for summarised dose-response data28The detailed steps of our literature search are shown in 2) and overall LC events. The SRR showed that being underweight, compared with the reference, was associated with an increased risk of LC  and the risk of LC were 0.82 (95%CI: 0.77–0.86) for overweight (Pheterogeneity = 0.02) and obesity (Pheterogeneity = 0.01). Therefore, we carried out subgroup analyses to identify sources of heterogeneity and minimise heterogeneity among eligible studies. As shown in The pooled RRs for the association between excess weight . As shown in Pnonlinearity < 0.01, I2 = 5%). Moreover, this dose-response trend showed a statistically significant decreased risk of developing LC with increasing BMI. Note that several representative point values are presented in For the dose-response analysis, data from 26 studies  were used, including 43,393 LC cases. The SRR and 95%CI of LC per five-unit increment in BMI were 0.97 (95%CI: 0.96–0.98) with significant heterogeneity . After stratifying by histologic type, there was a significant nonlinear association between BMI and lung adenocarcinoma risk  at 35 kg/m2, and 0.92 (95%CI: 0.72–1.17) at 40 kg/m2. For male non-smokers, the SRRs of LC risk were 0.96 (95%CI: 0.86–1.07) at the point of 30 kg/m2 and 0.95(95%CI: 0.71–1.27) at 35 kg/m2. For female non-smokers, the SRRs of LC risk were 0.95 (95%CI: 0.89–1.02) at the point of 30 kg/m2 and 0.95 (95%CI: 0.76–1.10) at 35 kg/m2. A sensitivity analysis for non-smokers was also conducted. Excluding those that were not controlled for potential confounders, none of the point estimates were materially altered, indicating that our results were robust.For non-smokers, it is noteworthy that the SRRs of LC risk were 0.96 (95%CI: 0.91–1.01) at the point of 30 kg/mP = 0.04), and the between-study variance was reduced from 0.008 to 0.000. The heterogeneity explained by the number of cases was 98.70%. In addition, the visual inspection of the Begg’s funnel plot or Egger’s plot revealed symmetry .For studies of those that were overweight, we found no significant heterogeneity between studies when stratified by the number of cases or study quality scores . In multsymmetry , and the2 increment of BMI corresponded to a 3.3% decrease in the risk of LC. BMI was inversely nonlinearly associated with overall LC risk. The reduction affects men and women alike, and it was greater for obesity than for overweight, displaying a significant nonlinear dose-response relationship. Note that the subgroup results of the dose-response analysis should be interpreted with caution because only a few studies focused on the histologic subtypes of LC, and interestingly, significant linear relationships that might result from heterogeneity and residual confounding were observed in current smokers, ever-smokers and non-smokers, respectively.The present meta-analysis is the first to provide comprehensive and quantitative evidence of the dose-response association between BMI and LC risk. We found that lower BMI was associated with higher LC risk, and each 5 kg/m2)et al. reported that a 5 kg/m2 increase in BMI was inversely associated with LC risk, and the pooled RRs were 0.76 (95%CI: 0.70–0.83) for men and 0.80 (95%CI: 0.66–0.97) for womenOne of the previous meta-analyses showed a negative association of 0.79 (95%CI: 0.73–0.85) between the risk of LC and excess weight  A allele, which was proven to correlate with increased BMI, was also associated with a decreased risk of LC47The theoretical biological plausibility for the negative association between obesity and LC risk is as follows. (1) Possible plausibility exists through the effect of adipose tissue on storage, mobilisation, and metabolism of carcinogen-DNA adducts363941434445A notable strength of the present methodology is the trend estimation of the summarised dose-response data, not only offering uniform analysis of cohort studies with different BMI categories and analysis of studies across varied levels of BMI, but also offering greater power using the full spectrum of continuous BMI data. Several limitations in our review should be mentioned. (1) There are several methodological limitations  that are worth consideration. (2) Unmeasured and residual confounding variables might bias the results toward the exaggeration or underestimation of risk estimates. Because of the inability to fully adjust for confounders, the protective effect of increasing BMI on LC could be attributed to other favourable lifestyle factors. (3) Due to the limitation of eligible data, the subgroups based on smoking status were crudely classified into current smoking, ex-smoking and non-smoking subgroups, regardless of the smoking duration, cumulative smoking exposure, and smoking intensity. (4) High heterogeneity across studies was presented for the overweight–LC and obesity–LC links, which would throw some doubts on the reliability of the SRR estimates. (5) Only studies published in the electronic edition of the databases in the English and Chinese languages have been reviewed for pertinence and studies in other languages were omitted. Despite these limitations, the meta-analysis included a substantial number of cohort studies, which greatly enhanced the statistical power of the analysis and provided enough evidence for the authors to draw a reliable conclusion.In summary, our meta-analysis contributes to the literature by exploring the nonlinear dose-response relationship between BMI and the risk of LC. An inverse dose-response relationship is likely to be present for LC risk with increasing BMI. Being underweight was strongly associated with increased risk of LC. Excess weight provides a potential protective effect against LC. However, the results from dose-response analyses in non-smokers support the idea that BMI has a modifying effect on smoking-related LC risk. Studies with large-scale, longer follow-up times and studies that are well-designed are warranted to confirm these subgroup results.How to cite this article: Duan, P. et al. Body mass index and risk of lung cancer: Systematic review and dose-response meta-analysis. Sci. Rep.5, 16938; doi: 10.1038/srep16938 (2015).


Evidence supporting the associations between folate metabolizing gene polymorphisms and pancreatic cancer has been inconclusive. We examined their associations in a case-control study of Japanese subjects.MTHFR) gene, one polymorphism (rs1801394) in the 5-methyltetrahydrofolate-homocysteine methyltransferase reductase (MTRR) gene and one polymorphism (rs1805087) in the 5-methyltetrahydrofolate-homocysteine methyltransferase (MTR) gene. Genotyping was performed using Fluidigm SNPtype assays. Unconditional logistic regression methods were used to estimate odds ratios (ORs) and 95 % confidence intervals (CIs) for the associations between folate metabolizing gene variants and pancreatic cancer risk.Our case-control study involved 360 newly diagnosed pancreatic cancer cases and 400 frequency-matched, non-cancer control subjects. We genotyped four folate metabolizing gene polymorphisms, including two polymorphisms (rs1801133 and rs1801131) in the methylenetetrahydrofolate  and 0.99 (0.65-1.51), respectively. For rs1801131, individuals with the CC genotype had approximately 1.2-fold increased risk compared with those with the AA genotype, but the association was not statistically significant. In analyses stratified by smoking and drinking status, no significant associations were noted for C677T genotypes. No significant interactions were observed with smoking and drinking with respect to pancreatic cancer risk.Overall we did not observe a significant association between these four genotypes and pancreatic cancer risk. For rs1801133, compared with individuals with the CC genotype of MTHFR polymorphisms or other polymorphisms in the folate metabolizing pathway are associated with pancreatic cancer risk.Our data did not support the hypothesis that  MTHFR has been the most extensively studied [MTHFR irreversibly converts 5,10-methylenetetrahydrofolate to 5-methyltetrahydrofolate, the predominant form of folate in the circulation. Two variants in the MTHFR, namely C677T (rs1801133) and A1298C (rs1801131), have been the focus of most studies [MTHFR A1298C, the homozygous CC genotype has approximately 60 % of normal MTHFR activity [Pancreatic cancer is often diagnosed at an advanced stage and has the poorest 5-year survival rate of any cancer. There is no effective screening method to identify apparently healthy individuals who are at risk for pancreatic cancer. The etiology of pancreatic cancer remains largely unknown, with only cigarette smoking and long standing diabetes being established as risk factors . The rol studied , 12. MTH77T rs18033 and A177T rs18033 and A1MTHFR, MTR, or MTRR genes and pancreatic cancer risk, with conflicting results [MTHFR , MTR (rs1805087) or MTRR (rs1801394) genes [MTRR gene, which encodes enzymes responsible for DNA methylation, has been shown to be associated with the susceptibility to pancreatic cancer in one previous Japanese case-control study [Evidence supporting the associations between folate metabolizing gene polymorphisms and pancreatic cancer has been inconclusive. To date, five case-control studies have examined the association between the polymorphisms in  results –17. Of t4) genes , 17. HowMTHFR C667T and A1298C genotypes resulting in decreased enzyme activity are associated with an increased risk for pancreatic cancer. Additionally, we performed analyses stratified by cigarette smoking and alcohol drinking to address the possibility that these two lifestyle factors may modify the associations between genetic polymorphisms and pancreatic cancer risk.Given that substantial evidence from epidemiologic and laboratory research supports an important role of folate in carcinogenesis , we genoThis case-control study was designed to address the role of genetic variations in determining pancreatic cancer risk in Japanese subjects. A detailed description of the study method has been published elsewhere . In brieFor the main effect of genotypes, the sample size was estimated using the following assumptions: multiplicative genetic model, 10 % minor allele frequency, 0.1 % disease prevalence, 1:1 case-control ratio, and OR = 2.0. Then the number of cases required to achieve 90 % power at a significance threshold of α = 0.05 was 190 cases . This stA self-administered questionnaire was used to solicit detailed information on demographic characteristics, medical history, and lifestyle factors such as smoking, drinking and dietary habits. Dietary habits were surveyed using a validated food frequency questionnaire (FFQ), in which the study subjects were asked to describe the usual intake frequency of 36 foods during the previous year prior to the study entry . After wThe genotyping of folate-metabolizing gene polymorphisms was performed using Fluidigm 192.24 Dynamic Array with BioMark HD Systems and EP1 . SNPtype assay , which employs allele-specifically designed fluorescences (FAM or VIC) primers and a common reverse primer, was used in this study. Genotype calls were obtained using the BioMark SNP Genotyping Analysis software. This software defined the genotype of each sample based on the relative intensities of fluorescences. The laboratory staff members were blinded to case or control status. Four quality control samples  were included in each assay, and the successful genotyping rate was 100 %.The differences in the characteristics of case patients and control subjects were tested using t-test, Mann-Whitney test, or chi-square test. The amount of daily ethanol intake was calculated based on the frequency and amount of alcohol beverages reported by the study subjects. The amount of daily dietary folate intake was estimated based on the responses to FFQ. A chi-square test was used to assess the Hardy-Weinberg equilibrium (HWE) in control subjects. Because the biological function of most SNPs has not been clearly defined, a co-dominant genomic model was assumed for SNP effects. Unconditional logistic regression models were used to estimate odds ratios (ORs) and 95 % confidence intervals (CIs) for the associations between folate metabolism-associated genetic polymorphisms and pancreatic cancer risk. All analyses were adjusted for age (continuous), sex , BMI , and cigarette smoking . All tests were two-tailed; a P value less than 0.05 was used to define statistical significance. The interaction of genotype and smoking and drinking with regard to pancreatic cancer risk was assessed using a likelihood ratio test. All statistical analyses were performed using SAS 9.12 .The distribution of genotypes for all SNPs among control subjects did not deviate from HWE. The selected characteristics of cases and controls are presented in Table MTHFR , MTRR (rs1801394), and MTR (rs1805087). Overall no significant associations were noted between any single genotype and risk of pancreatic cancer. For rs1801133, compared with individuals with the CC genotype of MTHFR C677T, the OR for those with the CT genotype and TT genotype was 0.87 (0.62–1.22) and 0.99 (0.65–1.51), respectively. For rs1801131, individuals with the CC genotype had approximately 1.2-fold increased risk compared with those with the AA genotype, but the association was not statistically significant.Table Table MTHFR genotypes on pancreatic cancer risk. No significant interactions were observed for C667T genotypes and cigarette smoking; the OR of pancreatic cancer was 1.94 (1.05–3.57) for individuals who had the TT genotype and were ever smokers. Similarly, no significant interactions were noted for C677T genotypes and alcohol drinking; the OR of pancreatic cancer was 1.05 (95 % CI: 0.62-1.78) for individuals who had the TT genotype and were ever smokers.Table MTHFR C677T and A1298C, and the risk of pancreatic cancer. Previous studies on the associations of MTHFR (C677T) with pancreatic cancer risk have shown inconsistent results [MTHFR C677T and A1298C genotypes [MTHFR C677T and A1298C genotypes [MTRR gene showed significant associations after multiple testing [MTRR gene, but further studies needs to replicate their positive finding.In this genetic case-control association study, we found no statistically significant associations between polymorphisms in folate metabolizing gene pathways, including  results –16. A ho results . A stron results . By contenotypes , 17. Ourenotypes . Furtherenotypes . HoweverMTHFR C677T, a key enzyme in the folate metabolizing pathway, we hypothesized that the TT genotype carriers have an increased risk of pancreatic cancer. Our results, however, showed that case patients had a similar distribution of the TT genotypes with that of control subjects, and neither CT nor TT genotypes were significantly associated with the risk when compared with CC genotypes. The reason for our null findings on the associations between MTHFR genotypes and pancreatic cancer risk is not clear, but the differences observed in minor allele frequencies among ethnic groups may in part account for the conflicting results. The genotype frequencies of MTHFR C677T differed between our case-control study and the previous case-control study that involved a US population [MTHFR 677C → T polymorphism is only associated with increased coronary heart disease risk in a setting of low folate intake [MTHFR. On the other hand, if folate intake is adequate, the 677 T variant of MTHFR preferentially routes one-carbon units to DNA synthesis at the expense of methionine, which is involved in DNA methylation [Given the lower enzymatic activity in individuals with the variant TT genotype of pulation . Even the intake . This fie intake . This mehylation . To addrMTHFR C677T and pancreatic cancer risk are smoking and drinking. Because alcohol is known as a folate antagonist and because smoking may impair folate status, several studies have evaluated the joint effects of the MTHFR polymorphisms with cigarette smoking or alcohol consumption in relation to pancreatic cancer risk [MTHFR 677 CC genotype [In addition to dietary folate intake, other lifestyle factors that may modify the associations between cer risk –15. In ogenotype , 14. In MTHFR genotypes, plasma levels and pancreatic cancer risk. It has been shown that DNA methylation was affected by genotype among only those with lower plasma folate levels [MTHFR genotypes, dietary folate, circulating folate level, alcohol, and/or other relevant nutrients including vitamin B2 and B12. Further studies need to develop a novel analytical method addressing the complex gene-environment interaction pathways involved in folate-derived carcinogenesis.Our study has several limitations. First, we had sufficient number of cases required to detect significant associations for the main effect of genotypes based on sample size estimation; however, we did not have sufficient power to detect significant associations in the analyses examining the synergistic effect of genotypes and exposure of interest. Assuming a dominant genetic model, a dichotomous exposure prevalence of 10 %, a relative risk for a genotype of 1.5, a relative risk for exposure of 1.5, and 1:1 case-control ratio, we need thousands of cases and controls to detect multiplicative interactions with relative risk of 2.0 . Second,e levels . MoreoveIn conclusion, our case-control study suggests that there is no association between genetic polymorphisms in folate-metabolizing genes and the risk of pancreatic cancer in Japanese subjects. The results may not be generalizable in view of limited sample size. Our study is also limited to address effect modification by alcohol, smoking or folate intake (gene-environment interaction) because of inadequate statistical power.MTHFR, methylenetetrahydrofolate; MTR, 5-methyltetrahydrofolate-homocysteine methyltransferase; MTRR, 5-methyltetrahydrofolate-homocysteine methyltransferase reductase; ORs, odds ratios.BMI, body mass index; CI, confidence interval; FFQ, food frequency questionnaire; HWE, Hardy-Weinberg equilibrium;


Family studies and twin studies demonstrate that lower urinary tract symptoms and pelvic organ prolapse are heritable. This review aimed to identify genetic polymorphisms tested for an association with lower urinary tract symptoms or prolapse, and to assess the strength, consistency, and risk of bias among reported associations.PubMed and HuGE Navigator were searched up to May 1, 2014, using a combination of genetic and phenotype key words, including “nocturia,” “incontinence,” “overactive bladder,” “prolapse,” and “enuresis.” Major genetics, urology, and gynecology conference abstracts were searched from 2005 through 2013. We screened 889 abstracts, and retrieved 78 full texts. In all, 27 published and 7 unpublished studies provided data on polymorphisms in or near 32 different genes. Fixed and random effects metaanalyses were conducted using codominant models of inheritance. We assessed the credibility of pooled associations using the interim Venice criteria.ADRB3 gene was associated with overactive bladder . The rs1800012 polymorphism of the COL1A1 gene was associated with prolapse  and stress urinary incontinence . Other metaanalyses, including those for polymorphisms of COL3A1,LAMC1,MMP1,MMP3, and MMP9 did not show significant effects. Many studies were at high risk of bias from genotyping error or population stratification.In pooled analysis, the rs4994 polymorphism of the ADRB3 with overactive bladder, and variation of COL1A1 with prolapse. Clinical testing for any of these polymorphisms cannot be recommended based on current evidence.These metaanalyses provide moderate epidemiological credibility for associations of variation in  Female pelvic floor disorders, an umbrella term including urinary incontinence, bladder storage symptoms, and pelvic organ prolapse (POP) are highly prevalent.The existence of inherited risk factors for pelvic floor disorders has been recognized for more than 150 years,Family studies provide limited information on heritability, as they do not control for shared exposure to environmental risk factors. Twin studies have been used to formally quantify the heritability of lower urinary tract symptoms (LUTS) or prolapse. In a sample of 16,886 Swedish twins aged >50 years, heritability was estimated as 41% for stress incontinence surgery, and 43% for prolapse surgery.Identification of the genetic variants underlying the heritability of these conditions would provide useful markers for clinical risk, prognosis, and treatment response. In addition, however, the insights provided should help explain the pathogenesis of these complex diseases, potentially offering new drug targets and preventative strategies. The aim of this systematic review was therefore to assess which candidate polymorphisms and/or candidate genes had been tested for an association with POP or LUTS in women, and to assess the strength, consistency, and potential for bias among published associations.The review protocol was prospectively registered (PROSPERO 2011:CRD42012001983).There are no gold standard diagnostic methods for either stress urinary incontinence (SUI) or other LUTS, as these are largely subjective symptomatic diagnoses. For POP, validated staging systems, including POP Quantification, have been widely used, but again there is no universally accepted criterion for diagnosis. We therefore expected to accept diagnostic criteria for LUTS and prolapse as specified within each study. In view of heterogeneity in definitions across studies, we tested for heterogeneity between studies with different criteria in different settings. We accepted definitions based on symptom questionnaires, clinical examination, urodynamics, or other validated assessments. We considered the population of interest as women aged ≥18 years.(polymorphism OR SNP OR CNV OR “copy number variation” OR mutation OR genetic OR chromosome OR VNTR OR InDel OR microsatellite) AND (nocturia OR LUTS OR incontinence OR urgency OR “overactive bladder” OR prolapse OR “Lower Urinary Tract Symptoms”[Mesh] OR “Urinary Incontinence”[MeSH] OR “enuresis”[Mesh] OR “Pelvic Organ Prolapse”[MeSH]) NOT mitral NOT carcinoma[Title] NOT cancer[Title] NOT .We combined searches from PubMed, HuGE Navigator, and an extensive selection of genetic, urological, and urogynecological conference reports. We searched PubMed up to May 1, 2014, without language restrictions, using a combination of genetic and phenotype key words and Medical Subject Headings (MeSH) terms: (“urination disorders” OR “urinary incontinence” OR “pelvic organ prolapse”).We searched HuGE Navigator, also through to May 1, 2014, using the following phenotype indexing terms: In addition we searched conference abstracts for annual meetings of the American Society of Human Genetics, American Urological Association, American Urogynecologic Society, European Association of Urology, European Society of Human Genetics, International Continence Society, International Urogynecological Association, and Society of Gynecologic Surgeons 2005 through 2013.We developed standardized data forms for this study, and conducted pilot screening and data extraction training exercises to achieve a high level of consensus between reviewers. All screening and data extraction was then performed independently and in duplicate by methodologically trained reviewers. Reviewers screened study reports by first screening titles and abstracts to select papers for full-text assessment, then screening full-text papers to confirm eligibility of the articles. Screening discrepancies were resolved by adjudication. We hand searched reference lists of all included articles, applying the same standardized screening process. When >1 report was identified for the same association in the same study population, we included the publication with the largest sample size.We contacted study authors by email, with a reminder after 1 month, for clarifications, additional information about methodology, and additional subgroup analyses where necessary. Data extracted included information on the setting for each study, details of the sampling strategy and sampled populations , the overall sample size and proportion genotyped, the outcome assessments used and phenotypic definitions, the genotyping method employed, and the genotyping quality control applied. Where possible we extracted or requested from authors full genotype frequencies among both cases and controls.2 statistic as a measure of between study heterogeneity. We recalculated the power of each study, and retested for departure from Hardy-Weinberg equilibrium. We made assessments of risk of bias in phenotype definitions, genotyping, and population stratification. We used the Harbord test of funnel plot asymmetry, and the significance chasing bias testFor polymorphisms assessed in ≥2 studies for the same phenotype assessed with similar case definitions, we conducted fixed or random effects metaanalyses as appropriate using the MetanWe screened 889 abstracts, and retrieved 78 full texts . In all,ADRB3); collagen, type I, alpha 1 (COL1A1); collagen, type 3, alpha 1 (COL3A1); laminin gamma 1 (LAMC1); matrix metalloproteinase-1 (MMP1); matrix metalloproteinase-3 (MMP3); and matrix metalloproteinase-9 (MMP9).Quantitative syntheses were possible for 11 polymorphisms in or near 7 genes: beta 3 adrenoceptor . There was significant deviation from Hardy-Weinberg equilibrium in one sample,Two studies of PolishA large number of mutations in collagen, type 3, alpha 1 have been associated with vascular Ehlers-Danlos syndrome. Inconsistent evidence suggests that urinary incontinence and prolapse may be prevalent among women with Ehlers-Danlos.2 = 83.7%, P < .01). Case definitions were similar for the 2 studies, making this an unlikely source of heterogeneity. The primary Korean study had suggested a large protective effect of the minor allele, and the heterogeneity between studies might instead be explained by differences in populations, or a simple Proteus effect.For rs111929073, separate KoreanLAMC1 was initially proposed as a candidate gene for prolapse in a linkage study of 9 individuals from a family affected by early-onset severe prolapse.Laminin gamma 1 is 1 of 3 kinds of laminin chain that combine to make different laminin isoforms. These extracellular matrix glycoproteins are an important constituent of basement membranes, with roles in cell adhesion and migration. All 3 individual studies found no association for rs10911193, with a nonsignificant pooled effect   and no hMMP1 gene is up-regulated in pelvic tissues of women with prolapse.Matrix metalloproteinase-1, also known as interstitial collagenase, is one of a number of enzymes that cleave collagen type 1. The MMP1, its common variants have received most research attention in association with cardiovascular disease,Matrix metalloproteinase-3, also known as stromelysin-1, is an enzyme that degrades a number of extracellular matrix components including collagen type 3 and elastin. Similarly to MMP1 and MMP3, its common polymorphisms have been linked to chronic obstructive pulmonary disease,2 = 68.9%, P = .04). Case definitions were similar for the 3 studies, making this an unlikely source of heterogeneity. All studies demonstrated Hardy-Weinberg equilibrium, and we judged a low risk of population stratification. The single study among Asian womenMatrix metalloproteinase-9, also known as 92-kDa type IV collagenase, degrades collagen type 4 and type 5. Some evidence suggests increased activation of MMP9 in pelvic tissues from women with prolapse.P > .1). We applied the significance chasing bias test,P < .1 is usually taken as the threshold for significance. We applied the test across each of the 13 metaanalyses conducted individually, and for the 13 considered together as 1 domain. Given the power of the individual studies to detect the observed pooled effect sizes in each metaanalysis, across the domain as a whole we expected 6.61 statistically significant studies, and observed 7 significant studies in our own prespecified reanalyses using the allelic test (P = .87). However, primary publications applied a variety of analytic techniques, and from the set of studies included in metaanalysis we observed 11 studies reporting statistically significant results in their own analyses (P = .14), typically using alternative models of inheritance. These findings are suggestive primarily of selective analysis, rather than publication bias. Individual metaanalyses again provided limited power for this test, but possible bias was most apparent in the quantitative synthesis of association of prolapse with the rs17576 SNP of MMP9 (P = .11).Each metaanalysis included at most 4 studies or subgroups, providing low power for conventional measures of funnel plot asymmetry. The Harbord test demonstrated no evidence of small study bias or publication bias GSTP1) and prolapse.P < 5 × 10-8)   in discoThe strengths of this review include a comprehensive search of both published and unpublished studies, applying explicit criteria to potentially eligible studies, and employing standardized, piloted data forms for data collection, guided by written instructions, and an unbiased assessment and synthesis of reported associations. We followed a prespecified data analysis plan, and contacted authors for clarifications and additional data.Among the challenges faced in this review was the inclusion of studies with varying diagnostic criteria. There may be considerable disparity between symptomatic and objective findings for both LUTS and prolapse, and despite long-standing efforts for standardizationIt is evident that overactive bladder in particular may have multiple underlying causes,Future advances are likely within the context of GWAS using large-scale population-based cohorts phenotyped for these conditions. The discovery of further causative variants should both help to explain the complex pathophysiology of these conditions, and provide potentially a route to effective prevention and treatment.ADRB3 with overactive bladder, and COL1A1 with prolapse. As for all complex diseases, these 2 currently identified polymorphisms explain a tiny fraction of that phenotypic variation. The widespread availability of direct-to-consumer testing means that some patients may present with questions about the implications of these polymorphisms. However, testing for any of these SNPs cannot be recommended based on current evidence. Nevertheless, clinicians and researchers should be aware of the putative risks associated with these SNPs, and the uncertainty regarding potential biases in the primary studies. In the future, genetic counseling may play one part of advice about risks of mode of delivery, and may help target women for primary or secondary prevention. Currently, clinicians should continue to use a family history of prolapse or incontinence as a simple marker of future risk, with clearly documented interactions with modifiable risk factors such as vaginal childbirth and obesity.Family and twin studies have provided convincing evidence for genetic predisposition to incontinence, prolapse, and overactive bladder, with genetic variation contributing up to half of population phenotypic variability. These metaanalyses provide moderate epidemiological credibility for associations of variation in


The purpose of this study is to investigate whether the Y402H polymorphism  in the complement factor H (CFH) gene have a pharmacogenetics effect on the anti-vascular endothelial growth factor (VEGF) treatment for neovascular age-related macular degeneration (AMD). We performed a meta-analysis using databases including PubMed and EMBASE to find relevant studies. 13 published association studies were selected for this meta-analysis, including 2704 patients. For the CFH Y402H polymorphism, anti-VEGF treatment was much less effective in AMD patients with the CFH CC genotype (CC versus TT: odds ratio (OR) = 55, 95% confidence interval (CI), 0.31 to 0.95, P = 0.03; CC versus CT: OR = 0.60, 95% CI, 0.40 to 0.91, P = 0.02; and CC versus CT + TT: OR = 0.59, 95% CI, 0.38 to 0.90, P = 0.02, respectively). In subgroup analysis, CFH Y402H polymorphism was more likely to be a predictor of response for Caucasians . In conclusion, pharmacogenetics of CFH Y402H polymorphism may play a role in response to anti-VEGF treatment for neovascular AMD, especially for Caucasians. Age-related macular degeneration (AMD) is the leading cause of irreversible blindness in people aged over 50 in the developed world348Genetic factors play an important role in the development of AMD11121314Although the role of CFH Y420H polymorphism in the overall risk for developing any form of AMD in general and neovascular AMD in particular is well-established, there is still some controversy about its role in the response to anti-VEGF treatment. Thus, patients with the CFH Y420H CC genotype had a lower visual acuity outcome in one studyTo the best of our knowledge, only one report conducted a meta-analysis focusing on the relationship between the CFH Y402H polymorphism and treatment response of neovascular AMD, indicating that CFH Y402H polymorphism might be associated with treatment response outcome in neovascular AMDA total of 658 articles were initially identified. Of these, 645 were rejected according to the exclusion criteria listed above. Hence, 13 studies were included in this meta-analysis181922232425262728293031We calculated a pooled OR based on genotype contrast. The results of meta-analysis for the CFH Y402H polymorphism and treatment response of neovascular AMD with anti-VEGF agents are summarized in In this analysis, the presence of CC versus TT genotype and CC versus CT genotype had a significant effect on the improved outcome as a result of anti-VEGF therapy , while the presence of the CT genotype versus TT did not have a significant effect (p = 0.65). The genetic model indicating mode of inheritance is most likely to be recessive, which compares the CC genotype with the combination of CT and TT genotypes. In the contrasts of the CC versus CT + TT model, the patients with the CC genotype appeared to be associated with a reduced response to anti-VEGF treatment for neovascular AMD . In this comparison heterogeneity was also identified and thus, a random-effects model was applied to the data . In a suOne limitation of the analysis when all trials were included is the heterogeneity of the positive treatment outcome, as shown in 18192223262728Pharmacogenetics examines the impact of genetic variation on the response to drugs. It has been suggested that genetic factors may influence response to anti-VEGF treatment in neovascular AMD8et al. found that patients with the CC genotype required had a 37% significantly higher risk of requiring additional ranibizumab injections over the first 9 monthsThe complement system, which is part of the immune system, plays an important role in inflammationBased on the current analysis, we propose that patients harboring the CFH CC genotype experience a reduced response to anti-VEGF treatment. When a positive response to treatment is defined more narrowly as whether or not an improvement in visual acuity occurs, the difference in response based on CC vs. CT/TT genotype becomes highly significant. Although the exact mechanism underlying an association between CFH genotype and anti-VEGF treatment response is currently unknown, it can be hypothesized that it may involve an enhanced inflammatory response due to an aberrant complement cascade behavior and an inferior capacity of the immune system to downregulate VEGF levels in the retina. As a result, patients with the homozygous CHF Y402H genotype (CC) and, to a much lesser extent, patients with the heterozygous genotype (CT) are likely to have a less favorable response to anti-VEHF treatment, and may require additional applications or switching to different anti-VEGF agents.Previous meta-analyses showed that the frequency of CFH Y402H C-allele is high in Caucasians (50~58%)This work may have some limitations. First, we cannot fully exclude publication bias. It is possible that some works, especially those published in languages other than English may have been missed. Second, a potential source of heterogeneity is different trial duration and different definition of improved anti-VEGF treatment response, the results should be interpreted with caution. Third, because of the complex nature of AMD, it is unlikely that a single nucleotide polymorphism in a single gene would be the only one associated with an increase in AMD risk and treatment response, without consideration of other polymorphic susceptible genes. For example, the ARMS2 and the VEGF-A genes could be harboring polymorphisms that can have an additional or separate effect on anti-VEGF treatment efficacy. To date, very few studies considered the cumulative effect of risk alleles in multiple genes  with inconsistent results9In conclusion, this is the first detailed meta-analysis to focus on the influence of genetic background on anti-VEGF treatment results. Our analysis provide evidence that pharmacogenetics of CFH Y402H polymorphism likely play a role in the frequency of the positive outcome to anti-VEGF treatment for neovascular AMD, especially in Caucasians and when improvement in visual acuity is defined as a primary outcome measure. Additional prospective studies with larger sample sizes would be helpful to confirm this association and study the influence of other genotypic variations to the treatment of choice.(“complement factor H” or “CFH”’) and (“age-related macular degeneration” or “AMD”). A manual search was performed by checking the reference lists of original reports and review articles to identify studies not yet included in the computerized databases. The final search was carried out on January 25, 2015, without restrictions regarding publication year or language.We conducted searches of PubMed and Embase, using the terms Articles were considered eligible for inclusion in the meta-analysis if the studies met the following inclusion criteria: (1) evaluating the relationship between the CFH Y402H polymorphism and the response to anti-VEGF treatment for neovascular AMD, (2) Independent retrospective or prospective association study, and (3) With sufficient available data to estimate an OR with 95% CI. Abstracts from conferences, full texts without raw data available for retrieval, duplicate publications, letters, and review articles were excluded.The data were extracted independently by two reviewers (G.H.C. and W.S.L.). Disagreement was resolved by discussion. The information extracted from each study included the authors of each study, the year of reported, information on study design, location and ethnicity of the trial, treatment modality, number of subjects, duration of the study, and genotype distributions.We assessed quality of included studies by a modified checklist based on the NOS2 statistic was calculated to assess heterogeneity between studies . If there was heterogeneity between studies, a random-effects model was applied to the data. Alternatively, a fixed-effects model was used for pooling the data. Genotype contrasts, including CC versus TT, CT versus TT, and CC versus CT were analyzed. The most appropriate genetic model was chosen for further analyzed. We also performed subgroup analysis by ethnicity. Harbord’s testThe quantitative data were entered into Cochrane Review Manager . The pooled OR with 95% CI was calculated by a fixed-effects model or a random-effects model according to the heterogeneity. P < 0.05 was considered statistically significant on the test for overall effect. The IHow to cite this article: Chen, G. et al. Pharmacogenetics of Complement Factor H Y402H Polymorphism and Treatment of Neovascular AMD with Anti-VEGF Agents: A Meta-Analysis. Sci. Rep.5, 14517; doi: 10.1038/srep14517 (2015).