Sinonasal cancer (SNC) has been related to occupational exposures, but the relative risk associated to specific jobs and/or carcinogen exposures other than wood and leather dust is generally based on small or inadequate sample sizes and the range of observed estimates is large. This paper is aimed at investigating such relationship through a systematic review of the literature followed by a meta-analysis of studies meeting specific inclusion criteria.2 index. When the heterogeneity hypothesis appeared relevant, the random-effect model was chosen. Sources of heterogeneity were explored using subgroup analyses.Systematic search was made with PubMed, Google Scholar and Scopus engines using related keywords. Occupational exposures include wood and leather dust, formaldehyde, nickel and chromium compounds, textile industry, farming and construction. Meta-analysis of published studies after 1985 with a case-control or cohort design was performed, firstly using the fixed-effect model. Heterogeneity was assessed with the Q statistical test and quantified by the Ipooled = 5.91, 95% CI: 4.31-8.11 for the case-control studies and 1.61, 95% CI: 1.10-2.37 for the cohort studies), as well as to leather dust . The strongest associations are with adenocarcinomas . An increased risk of SNC for exposures to formaldehyde , textile industry , construction  and nickel and chromium compounds  was found. Subset analyses identified several sources of heterogeneity and an exposure-response relationship was suggested for wood dust (p = 0.001).Out of 63 reviewed articles, 28  were used in the meta-analysis. Heterogeneity among studies was observed and random-effects models were used. Exposure to wood dust results associated with SNC . Future studies could be focused on specific occupational groups to confirm causative agents and to define appropriate preventive measures.The online version of this article (doi:10.1186/s12885-015-1042-2) contains supplementary material, which is available to authorized users. Sinonasal malignant neoplasms  are rare tumors with annual incidence rates around 1 per 100,000 in most developed countries. They represent less than 1% of all neoplasms and less than 4% of those arising in the head and neck region , while iA recent US analysis on sinonasal cancer (SNC) incidence and survival found that almost half of SNCs are localized to the nasal cavity 43.9%), most others originated in the maxillary 35.9%) or ethmoid (9.5%) sinus. These lesions were composed mostly of tumors of epithelial origin, including squamous cell carcinomas (SCC: 51.6%), adenocarcinomas (AC: 12.6%), esthesioneuroblastoma (ENB: 6.3%), and adenoid cystic carcinoma (ACC: 6.2%) [.9% or et%, most o3). Lower concentrations have been detected in finishing departments of plywood and particle-board mills and in the workroom air of sawmills and planer mills near chippers, saws, and planers. On the basis of the marked increase in the occurrence of SNC among woodworkers, the IARC concluded to assign wood dust to the group 1 of carcinogens to humans [A number of substances and occupational circumstances causing or possibly causing SNC have been classified by IARC as Group 1 and 2A. Tobacco smoking and other life-style factors seem to play a minor role (if any) in the etiology of SNC ,11. Occuo humans . SinonasHigh relative risks of SNC have been observed also for specific chemical exposures and occupational settings, including textile industry ,17,25-28Several other occupations were found associated to SNC in case-control studies but most of them lack statistical power to identify excess risks in specific jobs .A relationship between histotypes, anatomical site and occupational exposures has not been established clearly because only in few studies the onset site of the disease was exactly determined . NonetheThrough a systematic review followed by a meta-analysis, this study is aimed at investigating the possible relationships between occupational exposures and SNC risk in view of suggesting opportunities for prevention.http://www.ncbi.nlm.nih.gov/sites/entrez), Google Scholar (http://scholar.google.it/) and Scopus (http://www.scopus.com/) using as key words “sinonasal”, “cancer”, “occupational”, “risk”, “epidemiology”. We looked for additional studies by checking references in all identified publications. Complete articles about occupational risks for SNC were used to collect the following information: publication year (from 1968 to 2013), time period, type of publication, language, study design, topic, population studied, anatomical site, histologic subgroups, carcinogen exposure limits and criteria used to evaluate the quality of the evidence .A systematic review of studies was performed by a qualitative summary of published results. We conducted a search of PubMed ;Including the SNC subtypes AC and SCC;Referring to occupations and/or occupational setting with a potential risk of SNC;Exposure or potential exposure to specific risk factors stated explicitly, or from an industry/economic-activity recognized as having exposure to the risk factor ;Providing effect estimates with the corresponding measures of variability, or available data allowing for their calculation.Subsequently, studies were included in the meta-analysis when they complied with the following inclusion criteria:Finally, studies were excluded if they did not report original results  or did not provide sufficient data (e.g. lack of information about the number of cases and controls or about the used method).An abstract form of the most relevant available information  was created. When selected articles provided risk measures (OR/RR/SMR/SIR) stratified by specific variables , all the reported estimates were taken into account. An overview of the characteristics of the included studies can be found in Table Exposures were classified according to the HSE method  that is Separated analyses were performed for case-control and cohort studies, for all SNC grouped together and for each group of exposures. For case-control studies separate pooled risk estimates were calculated by the most common subtypes . Other histologic types were not studied because the classifications used were not comparable between studies and no information on occupational exposures were available. When not stated, crude risk estimates and 95% confidence intervals were calculated with the reported numbers, standard deviations were calculated by their confidence intervals  and the number of cases or controls, exposed or not exposed, was calculated starting from the risk estimate and the sample size.In order to calculate the pooled estimate and its confidence intervals, we first used a fixed-effect model with the inverse variance weighting method  with pooTi is the log risk ratio for the ith study and its weight ŵi is the inverse of variance (1/Si). Confidence intervals were obtained by normal approximation. When the heterogeneity hypothesis appeared relevant, the random-effect model with DerSimonian and Laird estimation method was preferred [where referred .The amount of variation between the collected effect sizes is shown together with the pooled estimates by the forest plots.Q statistical test was used to determine the homogeneity among the studies, with degrees of freedom equal to the number of studies minus one.The I2index was used to quantify the heterogeneity among studies as the percentage of the total variation not attributable to chance [o chance . The conWhen heterogeneity was present and data were able to be stratified (homogeneous strata among studies containing at least five estimates) potential sources of variability were explored through subgroup analyses. For formaldehyde we investigated the effect of exposure level  while for textile, wood and leather dust, that of exposure time . Although if for leather dust original exposure durations of selected studies were <15 and ≥12 years, according to the other analyses we consider as categories <15 and ≥15 years. Subgroup specific pooled relative risks were assessed and their difference was tested by the Z test.In order to assess publication bias we used funnel plots and Egger’s test , or nasal cancer risk factors other than occupational (e.g. lifestyle) or not included in the present study . Additional exclusion criteria were: previous studies of the same author/authors (the most recent have been chosen), books or book chapters, languages other than English, and studies without a case-control or cohort design .Residual 63 articles were reviewed and 28 out of these  met the inclusion criteria  by AC and SCC histologic subtypes are reported Table . SubgrouIn the following sections the epidemiological evidence about associations between occupational exposures and risk of SNC are summarized and the pooled risk estimates presented.3, an excess risk was found for AC  [A case study among wood-workers and an analysis from the Swedish Family-Cancer Database evidenced a significant high risk for SNC in all wood-related occupations ,50. Aden3-12.20) . A signi3-12.20) ,55. Thes3-12.20) . A retro3-12.20) .A French case-control study found significantly elevated risk of AC among male cabinetmakers , carpenters, joiners  and wood-working machine operators . Risks associated with cabinetmakers, carpenters and joiners were also significantly elevated for the other histologic subgroups: significant excesses in risk of SCC were observed among carpenters and joiners having worked for at least 15 years in the wood manufacturing industry  . ElevateIn a case-control study in Piedmont , the occurrence of SNC was found significantly related to ever exposure to wood dust (OR = 11.4) with a risk for AC (OR = 58.6) 10-fold higher than for other histotypes. Furthermore the risk for AC doubled every 5-years of exposure period to wood dust (p < 0.0001), and significantly increased also for low intensity exposure . Despitepooled = 29.43, 95% CI: 16.46-52.61) supports the findings of previous studies [pooled = 1.46, 95% CI: 1.01-2.1). Heterogeneity among the pooled risk estimates by length of exposure time suggests a possible exposure-response relationship   for the case-control studies Table  and 1.61 studies ,28,55. AAmong leather workers an increased risk of SNC (with AC as predominant) was found in an Italian study, associated with shoemaking  . Furtherpooled = 35.26, 95% CI: 20.62-60.28)   and a strong association with AC was found  and evidenced an exposure-response gradient, although confounding by wood dust could be of concern in some of the included studies . AnotherThe estimated total  AF for SNC associated with occupational exposure to formaldehyde is 0.17% (95%CI = 0.10-0.45) .pooled, low = 1.38) and moderate/high level of exposure , with non significant difference (p = 0.664)  for the case control (Table 3) the risk for SNC was higher than that for lung cancer, with a dose-response gradient for both nickel oxide and soluble nickel [3 - soluble in water, hexavalent chromium) was investigated in a group of chrome platers in UK, with a significant excess of occurrences (p < 0.05) observed in several death causes, including SNC  [An increased risk of SNC was found in nickel refinery workers ; in workers with the highest level of nickel exposure  . Similarth work) . The totth work) .In the present meta-analysis six studies about Nickel/Chromium exposure met our inclusion criteria, contributing a total of 25 effect estimates. The pooled relative SNC risk was 18.0 (95% CI: 14.55-22.27)  and a moderate increase in risk of AC  was observed in a case-control study in France . An incrIn France, significant deaths excess of SCC were observed for construction workers  and for farm workers of both sexes  .pooled = 2.03, 95% CI: 1.47-2.8) which increased for AC , with significant heterogeneity .The pooled relative risk for exposure in the textile industry from all 29 risk estimates of the six selected articles in this study, evidenced a significant risk of all SNC , while farming resulted non significantly associated with the risk of SNC  neither with SCC .In the construction sector  a significant pooled risk has been observed  and occupational exposures, the incidence rate rise exponentially. Due to the biunivocal and strong relationship between incidence and exposure to specific carcinogens, SNC is recognized as an “occupational tumor” together with malignant mesothelioma, supporting the development of specific epidemiological surveillance systems .Results from our meta-analysis indicate that SNC is related to wood dust exposure with a substantial heterogeneity among individual study estimates. Summary estimates also reveal positive associations between SNC risk and exposures to leather dust, formaldehyde, nickel/chromium compounds and in textile and construction industries.Currently, review articles and meta-analyses for assessing small risks with large public interest or important implications for public health have been increasing, though the use of meta-analysis of observational epidemiological studies draws less consent than in the area of clinical trials. The main concern relates to the synthetic approach of meta-analysis that emphasizes summarizing evidence over the search for heterogeneity . In our A moderate/high level of exposure to formaldehyde shows a higher risk of SNC risk than lower levels of exposures, but not statistically significant. Analyses on farming exposure did not provide any significant risk for SNC (p > 0.05), while an increased risk for SCC was found in the construction sector. Evidence of publication bias was detected only for wood dust exposure, where the regression method indicated minor risks for smaller studies.Limitations of this meta-analysis derive from the specific characteristics of the included studies and from the general ones of the used analysis. Firstly, a greater number of studies and risk estimates have been published on wood dust exposures with respect to other exposures, thus the pooled risk estimates are affected by different sizes. Secondly, the etiologic role of the exposure to carcinogens in different histological subtypes is still disputed and grouping together all SNC types could have reduced the causal role of occupational exposure. Anyway, the most of studies included in this meta-analysis focuses on AC and SCC as the most occupation-related subtypes . This is why analyses have been performed on the two subtypes separately and, for a rare disease such as sinonasal cancer, this increased the power to identify risk factors or to confirm previously suggested associations. Thirdly, differences across genders were not investigated, because only 5 case-control and one cohort studies showed separate results. Fourthly, studies used different exposure assessment, implying possible misclassifications. Finally, we have introduced a classification bias by considering the duration of exposure “≥12 years” as “≥15 years” in the subgroup analysis of leather dust.Although many studies assessed attributable fraction of SNC risk as reducing ,80,81 soEfforts should be made to improve informative campaigns and periodical medical checking because the anatomical site and the long latency often lead to SNC diagnosis only in advanced phases of the disease, while first tumour stages have a far better prognosis .The carcinogenic potential of some occupational hazard has been definitely established in several well-conducted epidemiological studies ,38,85-87By supporting the previously reported associations between occupational exposures and SNC incidence rate , our results may provide clues to the etiology of SNC.Our overall summary risk estimates strongly suggest that exposure to wood and leather dusts, formaldehyde, in the textile industry and to nickel/chromium compounds, increases the risk of developing SNC. A strength of this meta-analysis is that, by providing pooled SNC risk estimates, we have focused on this type of occupational cancer, that was scarcely emphasized by previous studies, mainly due to the small numbers of cases in studies.The failure to recognize SNC as an occupational disease may imply inadequate knowledge of SNC - a serious form of cancer with high levels of mortality, that significantly compromises quality of life - and incomplete preventive measures. Therefore, greater awareness should be required in exploring the occupational etiology of SNC, in medical monitoring, in implementing new technical solutions and discussing occupational threshold exposure levels. Finally the need to implement specific epidemiological surveillance system for occupational cancers must be considered.


Prospective observational studies have shown inconsistent associations of dietary or circulating n-3 long-chain polyunsaturated fatty acids (LCPUFA) with risk of all-cause mortality. A meta-analysis was performed to evaluate the associations. Potentially eligible studies were identified by searching PubMed and EMBASE databases. The summary relative risks (RRs) with 95% confidence intervals (CIs) were calculated using the random-effects model. Eleven prospective studies involving 371 965 participants from general populations and 31 185 death events were included. The summary RR of all-cause mortality for high-versus-low n-3 LCPUFA intake was 0.91 (95% CI: 0.84–0.98). The summary RR for eicosapentaenoic acid (EPA) and docosahexaenoic acid (DHA) intake was 0.83 (95% CI: 0.75–0.92) and 0.81 (95% CI: 0.74–0.95), respectively. In the dose-response analysis, each 0.3 g/d increment in n-3 LCPUFA intake was associated with 6% lower risk of all-cause mortality ; and each 1% increment in the proportions of circulating EPA and DHA in total fatty acids in blood was associated with 20%  and 21%  decreased risk of all-cause mortality, respectively. Moderate to high heterogeneity was observed across our anlayses. Our findings suggest that both dietary and circulating LCPUFA are inversely associated with all-cause mortality. The potential benefits of marine derived n-3 long-chain polyunsaturated fatty acids (LCPUFA)  have received great attention for over 40 years4810Circulating concentrations of n-3 LCPUFA represent a good indicator of dietary intakes and endogenous metabolism131415161718192021A flow chart of study selection is reported in 232526272931321314151617181920211314151617192021The 7 studies on dietary n-3 LCPUFA were published between 2004 and 2015, covering 361 273 participants and 27 624 deaths during 5.0 to 24 years of follow-up. Dietary intakes were mostly assessed with self-administered food frequency questionnaires (FFQ), with the exception of one Japanese cohort14151617P-heterogeneity = 0.01, I2 = 62.9%) . Meta-regression analysis showed that the observed heterogeneity was not explained by pre-defined study and population characteristics . A sensitivity analysis conducted by omitting one study at each turn showed a RR range of 0.88 (95% CI: 0.81–0.95) to 0.93 (95% CI: 0.87–0.99), and the overall I2 reduced from 62.9% to 32.3% when the Chinese cohorts by Takata et al.13I2 = 51.5%) for EPA and 0.81  for DHA, respectively.A meta-analysis of the 7 prospective studies suggested a summary RR of 0.91 (95% CI: 0.84–0.98) for the highest compared with lowest categories of n-3 LCPUFA intake, with moderate heterogeneity (= 62.9%) . There wI2 = 55.1% and 38.3%, respectively) .Three of the 4 studies reported high-versus-low circulating EPA and DHA and risk of all-cause mortality. The summary RR was 0.74 (95% CI: 0.60–0.90) for EPA and 0.78 (95% CI: 0.64–0.93) for DHA, with moderate heterogeneity (ctively) . There wP-heterogeneity = 0.01, I2 = 70.2%) (P-nonlinearity = 0.004) (One study= 70.2%) . There w= 0.004) , with a I2 = 74.5% and 79.3%, respectively)  .All four studies were included in the dose-response analysis of circulating EPA and DHA and all-cause mortality. The summary RR was 0.80 (95% CI: 0.65–0.98) and 0.79 (95% CI: 0.63–0.99) for each 1% increment in the proportions of EPA and DHA in total circulating fatty acids, with moderate to high heterogeneity (ctively) . There wy >0.30) . There wIn this meta-analysis involving over 30 thousand deaths events from 11 prospective studies, both dietary and circulating n-3 LCPUFA are shown to be significantly associated with reduced risk of all-cause mortality, and the associations are similar for EPA and DHA.et al.2et al.et al.et al.In a recent meta-analysis of 12 prospective studies, Zhao The reasons for the disparate findings between observational and clinical studies are not fully understood, but several possibilities merit consideration. First, most of the RCTs, as argued by Mozaffarian and Wu5Third, the association of n-3 LCPUFA with health outcomes may be nonlinear. A 2006 pooled analysisFourth, it is possible that other nutrients  in fish or other foods rather than n-3 LCPUFA are beneficial. It is difficult, perhaps not possible for observational studies to accurately distinguish the effects between n-3 LCPUFA and these nutrients. Finally, even if n-3 LCPUFA are one of the causal components in fish, their consumption as part of a matrix of other nutrients in foods may be essential for the benefits. We consider the fourth possibility less likely because of multiple lines of evidence from experimental research, prospective observational studies, as well as human intervention trials supporting the potential cardiovascular benefits of n-3 LCPUFA.Major strengths of this meta-analysis include the prospective design of original studies and the large number of events involved in the analyses. However, several limitations of this study should also be acknowledged. Since this meta-analysis is based on observational studies, potential influences of residual or unmeasured confounders on our findings cannot be fully excluded. Dietary information was mostly collected with self-reported FFQs in the original studies, which may introduce measurement error and lead some participants to be misclassified. Such misclassification would likely be nondifferential in cohort studies and to attenuate any true association. This may partly explain the observation that the associations between circulating n-3 LCPUFA and all-cause mortality were stronger than those between dietary intakes and all-cause mortality. Furthermore, the common methods we used to detect potential publication bias may be of a limited power when the number of studies is relatively small. Thus, potential impacts of publication bias on our results cannot be completely excluded.In summary, this meta-analysis of prospective observational studies suggests that both dietary and circulating n-3 LCPUFA are significantly inversely associated with risk of all-cause mortality. More large prospective studies conducted among individuals with high intakes are needed to address whether there is a nonlinear association. Future well designed primary prevention trials that account for nutrition status, health conditions, and medication usages of participants are also warranted to confirm our findings and those from others.This study was planned, conducted, and reported in adherence to the guidelines of the ‘Meta-analysis Of Observational Studies in Epidemiology (MOOSE) group’1) the study design was prospective; 2) the exposure of interest was dietary or circulating n-3 LCPUFA; 3) the outcome of interest was all-cause mortality 4) relative risks (RRs) with corresponding 95% confidence intervals (CIs) were reported or could be estimated. When multiple publications from the same study were available, the one with the largest number of events was selected.Studies that met the following criteria were considered: Using a standardized data-collection form, the following data were extracted from each included study: the first author’s last name, publication year, country of origin, source of populations, study duration, age and sex of participants, number of events and participants, categories of n-3 LCPUFA, the maximally adjusted RRs with 95% CIs, methods for exposure assessments, and potential confounders accounted for in the statistical model. The study quality was evaluated with the 9-star Newcastle-Ottawa Scale (NOS)Q and I2 statisticsQ statistic, P < 0.1 was considered as statistically significant; and for the I2 statistic, the following cut-off points were used: <30% (little or no heterogeneity), 30–75% (moderate heterogeneity) and >75% (high heterogeneity). Potential publication bias was investigated with both Egger regression and Begg correlation tests44In a Japanese cohortet al.20et al.49P value for nonlinearity was calculated by testing the null hypothesis that the coefficient of the second spline is equal to zero. All statistical analyses were performed using STATA software, version 11.0 .Given the distinct cut-off points across studies, dose-response analyses were performed with the method proposed by Greenland and LongneckerHow to cite this article: Chen, G.-C. et al. N-3 long-chain polyunsaturated fatty acids and risk of all-cause mortality among general populations: a meta-analysis. Sci. Rep.6, 28165; doi: 10.1038/srep28165 (2016).


Prostate cancer is one of the most common noncutaneous malignancies in Western countries. Because there has been a debate regarding the relationship between the XRCC1-Arg399Gln and Arg280His polymorphisms and prostate cancer risk, we therefore performed this meta-analysis. The electronic databases PubMed, EMBASE, and Medline were searched prior to October 1, 2014. An odds ratio and 95% confidence interval were used to calculate association. Heterogeneity was tested by both a chi-square test and I Prostate cancer (PC) is one of the most common noncutaneous malignancies in Western countries, such as Australia, the United States and France. It is considered the second most common cause of cancer death in men2368910121314131416The X-ray repair cross-complementing group 1 (XRCC1) is one of the DNA repair genes. At least four DNA repair pathways operate on specific types of damaged DNA, including base excision repair (BER), nucleotide excision repair (NER), double-strand break repair and mismatched repair (MMR). BER removes small lesions, including oxidized or reduced bases, fragmented, alkylation and nonbulky adducts. Nucleotide excision repair removes larger lesions, which often result from environmental damage, including UV radiation and external carcinogens. MMR corrects replication errors (base-base or insertion-deletion mismatch) caused by DNA polymerase errors. Double-stranded DNA breaks are repaired through mechanisms involving the homologous recombination repair pathway323411161722232425262728We performed a comprehensive search of electronic databases, such as PUBMED, EMBASE, and Medline, using the following key words: ‘XRCC1’ OR ‘X-ray repair cross-complementing group 1,’ ‘polymorphism’ OR ‘variants,’ ‘prostate cancer or prostate carcinoma’ from their earliest available date to October 1, 2014. There were no language, sample size, time period, population, or type of report restrictions in the search. Two authors independently evaluated all associated publications to retrieve the eligible literature. In addition, the references in reviews or in eligible studies were checked to ensure that no relevant studies were missed in the database search. If a study reported a different sub-population, we treated it as a separate study in our meta-analysis24Studies that met the following inclusion criteria were considered eligible articles: (1) investigated the relationship between the Arg399Gln and Arg280His polymorphisms of XRCC1 and prostate cancer risk; (2) provided sufficient data, including the genotype allele frequency in cases and controls; (3) case control studies; and (4) for multiple publications reported from the same population, we chose the largest and most complete study. Accordingly, the exclusion criteria were as follows: (1) the study was not relevant to the XRCC1 polymorphism and prostate cancer risk; (2) the study reported unavailable data; (3) the study did not offer the source of controls in the article; and (4) case reports.Two authors independently extracted the data according to the inclusion and exclusion criteria listed above. We extracted the following information from each eligible study: the name of first author, the study country, source of control, type of genotype, and sample size (cases and controls). According to the source of controls, eligible articles were defined as hospital-based or population-based. The genotype distribution was assessed according to whether it followed the Hardy-Weinberg equilibrium (HWE) balance in the control group. To guarantee the accuracy of the information extracted from the study, a third investigator reviewed the final results. Discussions were conducted to reach an agreement when we had different views on the results.2-test-based Q-statistic, and the degree of heterogeneity was estimated with the IP < 0.10) or IP > 0.05 showed that the controls followed HWE balance. Potential publication bias was assessed using funnel plots and Egger’s test First, all statistical analyses were performed using the STATA program Version 12.0). For each study, the pooled odds ratio (OR) with the corresponding 95% confident interval (CI) were calculated to assess the strength of the relationship between the XRCC1-Arg399Gln polymorphism and risk of prostate cancer based on different genetic comparison models: a heterozygote model (AG vs. AA), dominant model (GG + GA vs. AA), recessive model (GG vs. AA + GA), and homozygote model (GG vs. AA). Accordingly, the relationship between theXRCC1-Arg280His polymorphism and prostate cancer risk was evaluated by the combined OR and its 95%CI based on four genetic comparison models: the heterozygote model (AH vs. AA), the dominant model (HH + HA vs. AA), the recessive model (HH vs. AA + HA), and the homozygote model (HH vs. AA). The heterogeneity among the studies was assessed by the χ.0. For eP < 0.05). Of the 14 case control studies, 5 were about the XRCC1-Arg280His polymorphism and susceptibility to prostate cancer; 4 were conducted in Asian populations; 7 were in Caucasian populations; and 5 were in African populations. Ten studies were population-based (PB), and the other 4 studies were hospital-based (HB). All studies were written in English. The main characteristics of the studies and the genotype distribution results of the HWE test in the control group are shown in A flow chart of the detailed selection procedure is shown in 2347111617232425262728P = 0.048)(P = 0.005)(P = 0.480) or the dominant model GG + GA vs. AA . With regard to the XRCC1-Arg280His polymorphism and prostate cancer, a significant association was found under the heterozygote model AH vs. AA (P = 0.001), and the dominant model HH + HA vs. AA (P = 0.213). Furthermore, in the subgroup analysis by ethnicity, we discovered that there was no significant association in Caucasian populations in any of the gene models (P > 0.05). A significant association between the XRCC1-Arg399Gln polymorphism and prostate cancer was observed in African and Asian populations under only the homozygote model and the recessive model. When stratifying the studies by source of control, we found an increased risk in the PB group under the recessive model GG vs. AA + AG . There was no significant association in the HB group in any of the gene models (P > 0.05). The detailed results are shown in  = 0.048). a and t = 0.005). b, wher = 0.012). c, the  = 0.004). d. HoweA sensitivity analysis was conducted to assess the stability of the overall results through the sequential omission of each individual study. In our meta-analysis, the results show that no single study could affect the overall results, which indicated the reliability of our results .P values from Egger’s test were greater than 0.05, suggesting no publication bias in our meta-analysis.The funnel plot and Egger’s test were conducted to assess the potential publication bias of the studies. The shapes of the funnel plots did not show any evidence of obvious asymmetry . SimilarA variety of exogenous and endogenous factors can constantly damage cellular DNA. Unrepaired damage can result in the formation of DNA lesions or apoptosis or may even lead to unregulated cell growth and cancer. Several responses may occur to prevent replication in the presence of genetic errors in vivo16333435This meta-analysis addressed the association between two polymorphisms in the XRCC1 gene and prostate cancer risk, involving 2124 cases and 2219 controls concerning the Arg280His polymorphism from 5 studies, a pooled total of 4215 patients with prostate cancer, and 4220 controls concerning the Arg399Gln polymorphism from 14 studies. The overall data showed that the XRCC1 Arg399Gln polymorphism may be a risk factor for prostate cancer under the homozygote model (GG vs. AA) and the recessive model (GG vs. AA + AG), indicating that individuals who have the Gln allele are more likely to have prostate cancer . In the subgroup analysis by ethnicity, we discovered that Arg399Gln had a significant correlation with prostate cancer in Asian  and African populations . A previous meta-analysis performed by Geng et al.Many studies have extensively investigated the XRCC1 Arg399Gln polymorphism and have reported that this polymorphism is related to different cancers in Asian populations, including gastric cancer4649There have been studies regarding the XRCC1 Arg280His polymorphism and other cancers in different populations. These studies have suggested that the XRCC1 280His polymorphism is associated with hepatocellular carcinoma (HCC) risk and bladder cancer in Asian populations5155Considerable heterogeneity was detected among the studies. We also performed a subgroup analysis according to the source of the control, but the heterogeneity could not be fully explained. The included studies might have sample selection bias, and the different characteristics of subjects and their ethnic backgrounds may have a certain influence. In addition, hospital-based controls might not be truly representative of the general populationAlthough we performed a comprehensive analysis, several potential limitations should be considered. First, the genotype distributions in the control group from one study did not follow HWE balanceIn summary, our meta-analysis evaluated all of the available published data, and the results suggested that a significant association between XRCC1-Arg399Gln and risk of prostate cancer was found under the homozygote model and the recessive model, particularly in Asian and Africa populations. A significant association between the XRCC1-Arg280His polymorphism and prostate cancer susceptibility was found under the heterozygote model and the dominant model. In view of some limitations of our meta-analysis, well-designed case control studies and larger population sizes are needed to validate the role of XRCC1 Arg399Gln and Arg280His polymorphisms in the development of prostate cancer.F.L., H.T. and Z.D. defined the research theme. J.Y., X.W. and W.Y. designed the methods, analyzed the data, prepared figures and tables. J.Y. wrote the main manuscript text. All authors reviewed the manuscript.


Background. Although the direct cause of chronic periodontitis is bacterial infection, the progression of this disease depends on genetic and environmental factors, and smoking is a known risk factor in the development and severity of the disease. An individual's susceptibility may be influenced by polymorphisms in the glutathione S-transferase genes. These genes encode enzymes that metabolize xenobiotic compounds. The aim of this study was to determine the frequency of GSTM1, GSTT1, and GSTP1 polymorphisms in Mexicans with chronic periodontitis. Methods. 60 Mexicans with chronic periodontitis (30 smokers and 30 nonsmokers) were studied. A peripheral blood sample was taken for subsequent DNA extraction. The genetic material was PCR-amplified followed by restriction fragment length polymorphism with the aim of identifying GST polymorphisms. Results. Polymorphisms in the GSTT1 and GSTP1 genes were not significantly different between the smokers and nonsmokers. However, there were significant differences (P = 0.05) between groups in polymorphisms in the GSTM1 gene. The patients with chronic periodontitis have a higher frequency of null and mutant polymorphisms in GSTM1, GSTT1, and GSTP1 compared with historical data from a healthy Mexican population. Conclusions. The presence of these polymorphisms may be a risk factor for the development of chronic periodontitis. According to the World Health Organization, periodontal diseases are the second most prevalent oral cavity diseases, trailing only dental caries. The latest oral health report from the World Health Organization April 2012) indicates that between 15 and 20% of the middle-aged adult population (35–45 years) has periodontal disease . Fo. FoμL inThe digestion products were verified by horizontal electrophoresis, using a 2% concentration of agarose. The identification of polymorphisms was based on the presence of DNA fragments of different sizes. For exon 5, 176 bp, 91 bp, and 85 bp fragments correspond to heterozygote P1b (a/b) while 91 bp and 85 bp fragments correspond to homozygote P1b (b/b) and a 176 bp fragment corresponds to wild type P1b (a/a). Regarding exon 6 polymorphisms, the 332 bp fragment corresponds to homozygote P1c (c/c); three fragments, 332 bp, 174 bp, and 158 bp, correspond to heterozygote P1c (a/c) and two fragments, 158 bp and 174 bp, correspond to wild type P1c (a/a) .To differentiate between the frequencies of each polymorphism the analysis performed was different in proportions tested due to the nature of the variables. Statistical analysis was performed using SPSS software support in the 20 version .The demographic and clinical characteristics of the subjects included in the study are presented in Regarding the characteristics of the disease found among smokers, there was an average loss of 1.8 more teeth compared to the nonsmokers. The results for the full mouth plaque score, calculus, and severity of the chronic periodontitis were very similar between groups.The results of the analysis of polymorphisms in GSTM1 and GSTT1 indicated that there was a higher incidence of the GSTM1 null allele of 66.7% in the nonsmokers group compared to 43.4% with the GSTM1 null allele in the smokers group. Most individuals in both groups had the GSTT1 null polymorphism; the percentages were 73% of nonsmokers and 56.7% of smokers. The results of the identification of GSTP1 polymorphisms were very similar. It was observed that the GSTP1b a/a genotype was only present in the smokers group with an incidence of 10%. No cases of the GSTP1c polymorphism were found in the homozygous form c/c in both groups .P ≤ 0.05)  .As no differences between the groups were found in the majority of the genes studied, the patients with chronic periodontitis were grouped regardless of whether they smoked or not. With such grouping, the data obtained show that for both GSTM1 and GSTT1, there is a higher incidence of null polymorphisms. The results of the combined genotype analysis show that 80% of the patients have a null allele, that is, combinations −/−, −/+, +/−, with the double null genotype being the most frequent, representing 40% . This coSimilarly, when analyzing the total population with chronic periodontitis, the data indicate that for polymorphism GSTP1b, only 5% of patients have the a/a genotype, while the remaining 95% have some sort of variation. In contrast, 81.6% of patients had the a/a genotype for the GSTP1c polymorphism. No patients with the genotype c/c were reported .The analysis of gene-environment interactions may be the best way to understand the increased susceptibility to chronic periodontitis. Inflammation is known to be a consequence of periodontal bacteria and harmful products derived from cigarette smoke. Additionally, GST is involved in the detoxification of toxins that cause oxidative stress; some studies suggest an association between chronic periodontitis and polymorphisms in genes encoding enzymes that metabolize compounds derived from tobacco smoke , 16, 17.Several authors have reported that smoking is a risk factor for periodontitis –27. We cBy combining the smoker and nonsmoker groups, we created a total sample of 60 Mexican subjects with chronic periodontitis. In this sample, 55% of patients had the GSTM1 null genotype, which is consistent with the data reported by Concolino et al. (2007). These authors found an increased presence of the GSTM1 null genotype in 73.9% of their study group comprising of Caucasians from central Italy with chronic periodontal disease, similar to our data. Both of these studies differ from studies in a Korean population conducted by Kim et al. (2004), who reported this allele in just 37% of their subjects. For the GSTT1 null genotype, our results showed a frequency of this allele in 65% in our patients, which differs from the Caucasian population data being 24.6% for this allele , 17.Montero et al. (2007) performed a research study to determine the frequencies of polymorphisms in GSTM1 and GSTT1 in a healthy Mexican population with parents and grandparents born in Mexico. These authors had a study population comprising of 150 individuals, and they found that GSTM1 null is present with a frequency of 45% and that GSTT1 null has a frequency of 9.7%. Pérez-Morales et al. (2008) performed a similar study comprising of 529 Mexicans, and these authors observed a frequency of 33.4% for GSTM1 null and 12% for GSTT1 null , 29.Comparing these data and our results using a population of 60 Mexicans with chronic periodontitis, an increase in the frequency of the GSTM1 null allele to 55% is observed and our data indicate that the GSTT1 null allele is present at a frequency of 65%. Moreover, there is also an increase in the frequency of the combined GSTM1/T1 double null genotype, which was present in 40% of our patients compared to 2.2% and 4.24% reported by Pérez and Montero, respectively. This is consistent with the hypothesis that the double null combination leads to a decreased capacity to detoxify, which could lead to an increased susceptibility to chronic periodontal disease.There are no data in any population showing the prevalence of GSTP1b and GSTP1c in relation to chronic periodontitis, but there are data on these genes in healthy populations. A study performed by Mejia et al. (2013) in 60 Mexican subjects reported frequencies for GSTP1b and GSTP1c polymorphisms. They reported that for the frequencies of the wild type P1b allele a/a was 33.3%; the frequency of the heterozygous P1b allele a/b was in 56.6%, and the frequency of the homozygous P1b allele b/b was 10% .Comparing both studies, we observed a decrease in the presence of the wild type allele a/a and an increase in the homozygous allele b/b in our group of patients. Additionally, it is notable that 95% of the sick population of patients had some polymorphism in this gene. With regard to the distribution of GSTP1c, Mejia et al. reported that 100% of the individuals in their study had the wild type genotype a/a , while iThe observed frequencies of GSTM1 and GSTT1 nulls, GSTP1b and GSTP1c mutants show that these genes may confer increased susceptibility to chronic periodontal disease.A better understanding of the function of the polymorphic genes GSTM1, GSTT1, and GSTP1 and a better understanding of the disease could contribute to the prevention of chronic periodontitis through personalized recommendations and specific intervention.In conclusion, within the limitations of this study, the results suggest that compared with historical data from a healthy Mexican population, the population with chronic periodontitis shows an increase in null and mutant polymorphisms in GSTM1, GSTT1, and GSTP1. Therefore, the presence of these polymorphisms may be a risk factor for the presence of chronic periodontitis.Importantly, the results of this pilot study comprise the first report of the presence of GSTM1, GSTT1, and GSTP1 polymorphisms in a Mexican population with chronic periodontitis. The future development of this research will be the realization of these tests in a larger group of patients to measure the relationship between susceptibility genes and chronic periodontitis.


P < 0.001, I2 = 88.2%). In summary, this meta-analysis supports that being fatherless is associated with a lower risk of prostate cancer. Because of the substantial heterogeneity and residual confounding, using other study designs to further explore this association and the underling mechanism is warranted.The previously reported association between fatherhood status and prostate cancer risk was controversial. We carried out the present meta-analysis of all relevant studies to summarize evidence on this association. A comprehensive literature search of studies was performed in PubMed, Web of Science, and the Chinese National Knowledge Infrastructure (CNKI) databases, covering all the papers published from their inception to September 2015. The combined risk estimates with 95% confidence intervals (CIs) were calculated using a random effects model. Heterogeneity and publication bias were also evaluated. A total of 11 studies were finally included in this meta-analysis. We found a significantly reduced risk of prostate cancer associated with being childless . There was statistically significant heterogeneity across the studies ( Prostate cancer is the second most common form of cancer in males worldwide, with an expected 1,111,700 new cases and 307,500 deaths in 20125et al.Recently, several researchers have explored whether fatherhood status is a potential risk factor for prostate cancer with conflicting results. A Swedish case–control study1415The detailed processes of literature search are presented in 9101114151617181920P < 0.001, I2 = 88.2%).The multivariable-adjusted ORs of childless men versus fathers, for each study and for the combination of all the studies, are presented in Then we conducted stratified analyses by study design, geographical region, number of cases, type of exposure, marital status, and education . In the Q test and the I2 index were adopted to examine the heterogeneity among included studies. As shown in P < 0.001 for heterogeneity, I2 = 88.2%). Then the Galbraith plot was used to detect the potential sources of heterogeneity. As shown in et al.et al.et al.P = 0.337 for heterogeneity, I2 = 12.0%) while the overall association remained significant   .In the sensitivity analysis, the influence of each study on the summary effect estimate was evaluated by repeating the meta-analysis after removing one study at a time. As shown in P = 0.533). However, Egger’s test suggested the existence of publication bias (P = 0.003). The trim-and-fill method identified one possible (imputed) missing study .Androgen status may mediate the link between fatherhood status and prostate cancer risk. Infertile men have lower levels of testosterone and serum testosterone to estradiol ratios compared with fertile men25It is worth noting that fatherhood status as a proxy for male fertility are hampered by the fact that the fatherhood status is influenced by various factors comprising fertility of the man, fertility of the partner, opportunity to start a family, and desire to have children. In subgroup analysis by marital status, the association was still statistically significant in the studies adjusted for marital status . Thus, the presence of confounding from unmarried men could be partially ruled out. However, we were not able to assess the differential misclassification from men’s reproductive intent and fertility of their partners.Another important confounding factor was the difference in healthcare-seeking patterns, as married and better educated men may have a higher uptake of PSA testingOur study had some strengths. A total of 182,012 prostate cancer cases were included in this meta-analysis, which enhanced the statistical power and provided more reliable estimates. The estimates from fully adjusted models in each study were used in this study to minimize potential confounding. Various subgroup analyses and sensitivity analyses were carried out to assess the robustness of the results.P < 0.001 for heterogeneity, I2 = 88.2%), which might distort the pooled estimates. Through the Galbraith plot, we detected the studies that potentially contributed to the heterogeneity. After removing these studies, the combined estimate remained significant  without obvious heterogeneity , which indicated that the heterogeneity didn’t have a material impact on our conclusion. Second, publication bias was detected by Egger’s test (P = 0.003), although the trim-and-fill analysis did not alter the findings substantially. Small negative studies were less likely to be published and we were not able to include some gray literature, such as conference abstracts and studies reported in languages other than English or Chinese. Third, half of the included studies adopted case-control design, which may introduce the possibility of select bias and recall bias. However, the observed association was robust and consistent in the subgroup of cohort/nested case-control studies . Fourth, there were vast differences in size of included studies, ranging from 168 to 117,328. The pooled estimate was vulnerable to the results of large studies, such as Giwercman et al.’s and Wiren et al.’s studies99Our study also had several important methodological limitations. First, substantial heterogeneity was observed across individual studies  databases, covering all the papers published from their inception to September 2015. We adopted the following search algorithm:  AND (prostate OR prostatic) AND  AND (cohort OR case-control OR case control). No language limitation was applied. We evaluated all retrieved publications carefully by examining their titles and abstracts, and the full texts of studies potentially matching the eligible criteria were further checked. We also reviewed reference lists of articles and reviews to identify any additional relevant studies. The present systematic review was planned, conducted, and reported in adherence to standards of quality for reporting meta-analysesi) the exposure of interest was being childless or infertility. Infertility is defined as the absence of a live birth for men who desire a child. Considering infertile men generally have no children, the studies with exposure of male infertility were also included in the pooled analysis; (ii) the outcome of interest was prostate cancer; (iii) study design was case-control or cohort; and (iv) the effect size estimates with their corresponding 95% confidence intervals (CIs) were provided . Whenever multiple articles from the same study population were available, the study with the largest number of cases and most applicable information was included in this meta-analysis. In particular, we included a pooled analysis of 18 studies, rather than separate articles, since part of these articles are non-English literature or lack sufficient data to calculate the effect size estimate.The studies included in this meta-analysis met all of the following criteria: . This scale has three categories: Selection (four items), Comparability (one item), and Exposure for case-control studies or Outcome for cohort studies (three items). A study can be awarded a maximum of one star for each numbered item within the Selection and Exposure/Outcome categories. A maximum of two stars can be given for Comparability. The stars are then added up to obtain the total quality score.The quality of each study was evaluated by two authors (Y.M. and X.X.) using the Newcastle-Ottawa Scale (NOS)  using a predesigned data collection form. Discrepancies were resolved by group consensus and consulting a third reviewer.Considering that prostate cancer is a rare disease, the relative risk (RR) was assumed to be nearly equal to the odds ratio (OR), and the OR was used as the study outcome. The combined OR with its 95% CI was calculated to assess the strength of the association between fatherhood status and prostate cancer riskQ statistic and the I2 scoreI2 was used to describe the degree of heterogeneity . Galbraith plotThe heterogeneity among included studies was tested by the Sensitivity analysis was conducted by repeating the meta-analysis after the omission of every study in turn to assess the effect of each study on the combined estimate. Cumulative meta-analysis was also performed by sorting the studies based on publication year.P < 0.05, results were considered statistically significant. All of the statistical analyses were conducted with STATA 11.0 , using two-sided P values.Potential small-study bias was evaluated by Begg’s test (rank correlation method)How to cite this article: Mao, Y. et al. Reduced risk of prostate cancer in childless men as compared to fathers: a systematic review and meta-analysis. Sci. Rep.6, 19210; doi: 10.1038/srep19210 (2016).


The associations were also significant both in Caucasian , Asian  and Mixed population . No publication bias was found. Our analysis suggested that the higher intake of vitamin C might reduce the risk of pancreatic cancer.Quantification of the association between the intake of vitamin C and risk of pancreatic cancer is still conflicting. We therefore conducted a meta-analysis to assess the association between them. Pertinent studies were identified by a search of PubMed and Web of Knowledge throughSeptember of 2014. A random effects model was used to combine the data for analysis. Sensitivity analysis and publication bias were conducted. Data from 17 studies including 4827 pancreatic cancer cases were used in this meta-analysis. Pooled results suggested that highest vitamin C intake amount versus lowest amount was significantlyassociated with reduced the risk of pancreatic cancer [summary relative risk (RR) = 0.705, 95% CI = 0.612–0.811, I Pancreatic cancer is the eighth most common cause of cancer death in Europe2Diet may be involved in the aetiology of pancreatic cancer and dietary variations between countries may explain the differences in incidence. For antioxidants, such as vitamins C, there are several plausible biological mechanisms by which they might prevent pancreatic cancer, including inactivating free radicals and reducing oxidative DNA damage, stimulating immune function111213141516171920212223242526Studies were identified using a literature search of PubMed and Web of Knowledge through September 2014 and by hand-searching the reference lists of the retrieved articles. The following search terms were used: ‘pancreatic cancer’ or ‘pancreatic carcinoma’ combined with ‘nutrition,’ ‘diet,’ ‘lifestyle,’ ‘vitamin C,’‘vitamins’ or ‘ascorbic acid’. Two investigators searched articles and reviewed all the retrieved studies independently. Disagreements between the two investigators were resolved by consensus with a third reviewer.For inclusion, studies had to fulfill the following criteria: (1) have a prospective or case-control or retrospective cohort studies; (2) vitamin C intake was the independent variable of interest; (3) the dependent variable of interest was pancreatic cancer; (4) relative risk (RR) or odds ratio (OR) or hazardratio (HR) with a 95% confidence interval (CI) was provided . If data were replicated in more than one study, we included the study with the largest number of cases. Accordingly, the following exclusion criteria were also used: (1) reviews; (2) repeated or overlapped publications.Two researchers independently extracted the following data from each study that met the criteria for inclusion: the first author’s last name, year of publication, geographic locations, study design, sample source, the age range of study participants, duration of follow-up, the number of cases and participants, and RR (95% CI) for vitamin C intake and pancreatic cancer risk. From each study, we extracted the RR that reflected the greatest degree of control for potential confounders. If there was disagreement between the two investigators about eligibility of the data, it was resolved by consensus with a third reviewer.To determine the quality score of included studies, two reviewersindependently performed the quality assessment by using the Newcastle-Ottawa Scale2 was used to assess heterogeneity, and I2 values of 0, 25, 50 and 75% represent no, low, moderate and high heterogeneityThe pooled measure was calculated as the inverse variance-weighted mean of the logarithm of RR with 95% CI, to assess the association between vitamin C intake and pancreatic cancer risk. Random-effects model was used to combine study-specific RR (95% CI), which considers both within-study and between-study variationThe search strategy identified 158 articles from PubMed and 243 from the Web of Knowledge, and 28 articles were reviewed in full after reviewing the title/abstract. In total, 17 articles111213141516171819202122232425262 = 42.3%] . Highest vitamin C intake level versus lowest level was significantly associated with the risk of pancreatic cancer both in Caucasian , Asian  and Mixed population . The detailed results are summarized in 2 = 42.3%, Pheterogeneity = 0.034) was found in the analysis. In order to explore the moderate to high between-study heterogeneity founded in several analysis, univariate meta-regression with the covariates of publication year, ethnicity, study design (case-control or prospective), number of cases and source of controls was performed. No significant findings were found in the above-mentioned analysis. Considering the adjustment of individual studies is heterogeneous, we then provide the original unadjusted relative risks and pooled them together to derive an effect size estimate. The pooled RR was 0.771 (95% CI = 0.685–0.868) for vitamin C intake and pancreatic cancer risk. Low heterogeneity was found .As shown in the pooled results, moderate heterogeneity (IP = 0.414) showed no evidence of significant publication bias between vitamin C intake and pancreatic cancer risk.Influence analysis showed that no individual study had excessive influence on the association of vitamin C intake and pancreatic cancer risk. The trim-and-fill funnels  and EggeFinding from this meta-analysis suggested that the higher intake of vitamin C could reduce the risk of pancreatic cancer. The associations were also found in subgroups of Caucasian, Asian and Mixed population for vitamin C intake and pancreatic cancer risk.Vitamin C is one of the most common antioxidants in fruits and vegetables, and it may exert chemopreventive effects2 = 42.3%, Pheterogeneity = 0.034) in our pooled results. This might havearisen from publication year, study design, geographic location, and sources of controls or number of cases. Thus, we used meta-regression to explore the causes of heterogeneity for covariates. However, no covariate having a significant impact on between-studyheterogeneity was found among those mentioned above. We then performed subgroup analyses by the type of study design (prospective or case-control studies) and ethnicityto explore the sourceof heterogeneity. However, between-study heterogeneity persisted insome of the subgroups, suggesting the presence of other unknown confounding factors. Considering the adjustment of individual studies is heterogeneous, we then provide the original unadjusted relative risks and pooled them together to derive an effect size estimate. The pooled RR was 0.771 (95% CI = 0.685–0.868) for vitamin C intake and pancreatic cancer risk. Low heterogeneity was found . This may be because the adjustment of individualstudies for confounding factors is different.Munafo and Flint reported that between-study heterogeneityis common in meta-analysesAs a meta-analysis of published studies, our findings showed some advantages. First, this is the first comprehensivemeta-analysis of vitamin C intake and pancreatic cancer risk based on highestamountversus lowest amount analysis. Second, large number of cases and participants was included, allowing a much greater possibility of reaching reasonable conclusions between vitamin C intake and pancreatic cancer risk. Third, no significant publication bias was found, indicating that our results are stable.P = 0.39) was found in the above-mentioned analysis. However, significant associations were found both in case-controls studies and in cohort studies. More studies with prospective design are wanted in the future studies while only 4 studies included in this meta-analysis were prospective design. Second, measurement errors are important in the assessment of dietary intake, which can lead to overestimation of the range of intake and underestimation of the magnitude of the relationship between dietary intake and cancer risk40P = 0.414) showed no evidence of significant publication bias was found suggesting that our results are stable. Fourth, there seems to be a big gap in published material between 1993 and 2002 in our meta-analysis. For this reason, we have searched the databasecarefully and by hand-searching the reference lists of the retrieved articles again, and did not find any related articles. Fifth, there appears to be a large variability in the baseline riskdue to the limitation of published material used, for whichwe cannot change. Finally, between-study heterogeneity was found in the pooled analysis, and the between-study heterogeneity was successfully explained by themeta-regression.There were also some limitations in this meta-analysis. First, a meta-analysis of observationalstudies is susceptible to potential bias inherent in the original studies, especially for case-control studies. Overstated association may be expected from the case-control studies because of recall or selection bias. Although case-control studycan allow a recall or selection bias, case-control studyis an important method in etiology research. In order to find whether the study design is a key contributor to the between-study heterogeneity, univariate meta-regression with study design (case-controlor cohort) was performed. No significant finding .


Fish and long-chain ω-3 polyunsaturated fatty acid (LCω3PUFA) intake in relation to the risk of cardiovascular diseases have been well studied. However, studies that directly link fish consumption or LCω3PUFA intake to the risk of metabolic syndrome (MetS) are sparse and the results are inconsistent. We reviewed literature through December 2014 and used random-effects or fixed-effects models, as appropriate, to pool the associations of fish or LCω3PUFA intake with the risk of MetS. Nine independent cross-sectional samples  and three independent prospective cohorts (two prospective cohort studies) were identified as eligible for this meta-analysis. By pooling data from the prospective cohorts (7860 participants and 1671 incident cases), a significant inverse association between fish consumption and incidence of MetS was found. The pooled RR (95% CI) was 0.71 , comparing the highest to the lowest category of fish consumption, and 0.94  for one serving/week increment. Consistent results were found for LCω3PUFA intake. Non-significant inverse association of fish or LCω3PUFA intake with risk of MetS was found when pooling the cross-sectional studies. By quantitatively summarizing the literature, a modest inverse association between fish or LCω3PUFA intake and risk of MetS cannot be excluded. Metabolic syndrome (MetS) is considered a major public health problem in the United States as its prevalence increased rapidly in the past two decades. The recent data from the National Health and Nutrition Examination Survey (NHANES) suggest that approximately one third of American adults suffers from this syndrome ,2. Theren-3 fatty acid”, “metabolic syndrome”, “metabolic syndrome X”, “syndrome X”, and “insulin resistance syndrome.” Additional information was retrieved through Google search and a hand search of the references of relevant articles.This meta-analysis was conducted according to MOOSE  guidelines in all stages of study design, implementation, and reporting . A systeWe proposed to search for original research articles including prospective cohort studies, case-control studies, and cross-sectional studies, which were published in English and provided hazards ratio (HR), relative risk (RR) or odds ratio (OR), and the corresponding 95% CIs of MetS in relation to fish or LCω3PUFA intake.in vitro or animal studies; (4) 14 were genetic studies; (5) 6 were not published in English; (6) 133 did not relate fish consumption or LCω3PUFA intake to the risk of MetS. Among the remaining 25 studies, 17 studies were further excluded because they did not provide the data in the form required for the meta-analysis. If a study reported results for male and female participants separately, the study was counted as two independent cohorts in the meta-analysis. An unpublished study with de novo results on fish and MetS from the Coronary Artery Risk Development in Young Adults (CARDIA) Study was also included in the present meta-analysis. In the final dataset for the meta-analysis, 7 cross-sectional studies (9 independent samples) and 2 prospective studies (3 independent cohorts) were included.As shown in The following data were extracted from the original publications: design of the study , the first author’s name, year of publication, study population, country of origin, size of sample, age range or mean age of the participants, proportion of men, duration of follow-up (for prospective cohort studies), method for diet assessment, categories of fish or LCω3PUFA intake, criteria for MetS ascertainment, adjusted covariates, and ORs or RRs with the corresponding 95% CI. All procedures, including the literature search, study selection, and data extraction, were independently performed by two authors (YSK and PX). Any disagreements were solved by group discussion.In this meta-analysis, OR  or RR (from prospective cohort studies) was used as a measure of effect size, and no eligible cohort studies reported HR. ORs or RRs and 95% CIs were transformed to their natural logarithms (ln) and were used to compute the corresponding standard errors (SEs). Since no significant heterogeneity was observed in most of the pooled analyses, results from the fixed-effects models were reported for the main analysis. The ln (ORs) or ln (RRs) from primary studies was weighted by the inverse of the within-study variance .We also determined the pooled dose-response relationship between fish (one serving/week increment) or LCω3PUFA intake (per 100 mg/day increment) and the risk of MetS in the prospective cohorts. If the primary study did not provide information on linear association, we estimated it based on the categorical analysis using the meta-analysis regression method . For insQ test was used to determine the heterogeneity among studies statistically [I2 was computed to quantify the degree of inconsistency across studies [I2 > 75% was considered as having strong heterogeneity across studies [Cochran’s stically . I2 was  studies ; a value studies . Publica studies .Sensitivity analyses included: (1) investigating the influence of a single study on the combined association by omitting one study at each time in the pooled analysis; and (2) examining whether the overall association was robust depending on modeling by replacing the fixed-effects model with the random-effects model in the pooled analysis.p ≤ 0.05 was considered statistically significant for all tests.All analyses were performed using STATA statistical software . A total of 9 independent cross-sectional samples and 3 independent prospective cohorts from 9 identified studies  were included in the meta-analysis . Sample I2: 60.7%, p = 0.08) and dose-response relationship analysis .In the pooled analyses of prospective cohorts, a significant inverse association between fish consumption and incidence of MetS was found comparing the highest to the lowest category of exposure . The incidence of MetS reduced by 6% for one serving/week increment in fish consumption . Marginally significant heterogeneity across studies was observed in both categorical analysis  .In addition, significant inverse association between LCω3PUFA intake and incidence of MetS was found comparing the highest to the lowest category of exposure  with a marginally significant heterogeneity across the included studies ( = 0.06) . The incI2: 0.0%, p = 0.74), whereas there was significant heterogeneity among studies for fish consumption . The pooled dose-response relationship in the cross-sectional studies was not assessed due to the insufficient data.In the pooled analyses of cross-sectional studies, statistically non-significant inverse associations between fish consumption or LCω3PUFA intake and risk of MetS were observed. The pooled ORs (95% CIs) comparing the highest to the lowest category of exposure were 0.85  for fish consumption and 0.94  for LCω3PUFA intake . No signp = 0.96 for the categorical analysis and p = 0.74 for the dose-response relationship assessment in the prospective cohort studies; p = 0.22 for the categorical analysis in the cross-sectional studies). Similarly, there was no evidence of publication bias for LCω3PUFA intake . Begg’s test confirmed these results.Egger’s regression asymmetry test indicated no evidence of publication bias for fish consumption .It has been hypothesized that fish or LCω3PUFA intake may be associated with lower risk of MetS . Howeveri.e., Kim et al. In addition, a few issues merit discussion. First, similar to any other meta-analysis, the inherent limitations of primary studies may have affected our findings. Second, although the individual OR or RR estimate in the primary studies was adjusted for different covariates, the possibility that unmeasured factors or residual confounding biased our findings cannot be completely excluded. For instance, the largest cross-sectional study [Our meta-analysis includes 3 prospective cohorts (~8000 participants) and 7 cross-sectional studies . While we acknowledge that the overall sample size may still not be sufficient , we believe that findings from this meta-analysis will generate valuable data and stimulate research in this field. Clearly, more longitudinal studies are needed in order to make any solid conclusion. In addition, the limited numbers of included studies may explain how the overall association from prospective cohort studies is influenced by one study, al study , which cal study . The inval study ,30, fishal study ,24,25,26al study . Of noteThe publication bias is always a concern in meta-analysis. Although we found little evidence of publication bias in the present meta-analysis, a potential publication bias resulting from excluding unpublished data or publications in non-English languages cannot be ruled out.The best approach to evaluate the causality of diet-disease relations is to conduct a long-term, double-blinded, and placebo-controlled randomized trial. However, given the practical and ethical limitations such as the participant’s long-term compliance, it may not be feasible to pursue such a trial on fish consumption and risk of MetS. Thus, we hope findings from this meta-analysis will draw researchers’ attention and call for longitudinal studies of fish consumption and risk of MetS.In conclusion, a modest inverse association between fish consumption or LCω3PUFA intake and risk of MetS has been observed when combining available data from prospective cohort studies, but not the cross-sectional studies. More research, especially prospective cohort study, is needed in order to make a solid conclusion and to investigate fish and LCω3PUFA intake in relation to individual components of MetS.


Estrogen receptors play an important role in mediating estrogen action on target tissues, and the estrogen is relevant to male infertility. Single nucleotide polymorphisms (SNPs) in estrogen receptors may be associated with the risk of male infertility. A variety of case control studies have been published evaluating this association. However, the accumulated studies have shown inconsistent conclusions.To further determine the potential association between the four common SNPs  in estrogen receptors gene and male infertility, this meta-analysis was performed according to the 10 published case control studies. The odds ratio (OR) and 95% confidence interval (CI) were used to evaluate the strength of the associations.It was revealed that the sub-group analysis by the ethnicity, for the rs2234693, a significant association in the comparison of CC vs. TT , CT vs. TT  and CC + CT vs. TT  in the Asian population with male infertility. For rs9340799 polymorphism, increased risks were observed for the comparison of AA vs. GG  and AA vs. GA + GG . For rs1256049 polymorphism, the comparison of the GA vs. GG  and AA + GA vs. GG , also increased risks present in Asian and Caucasian population, respectively.The rs2234693C allele was associated with the decreased risk for male infertility; however, the rs9340799AA genotype and the rs1256049GA genotype were associated with an increased risk for male infertility.The online version of this article (doi:10.1186/1477-7827-12-79) contains supplementary material, which is available to authorized users. Male infertility is an important cause of couple’s inability to bear children in 20% to 25% of total cases and the etiology of nearly half of the cases remains idiopathic, 2. ApprERs1 gene on chromosome 6q25, and ERβ is a 530-amino acid protein[ERs2 gene on chromosome 14q22-24[ERs have been identified to be two subtypes of ERα and ERβ. ERα is a 595-amino acid protein encoded  protein which en14q22-24. Genetic14q22-24, and two14q22-24. To date14q22-24, 19–26. To identify all articles that examined the association of ERs polymorphisms with male infertility, a comprehensive systematic bibliographic search through the medical databases PUBMED, attempting to cover all medical papers published between 1950 and 2013, using the following keywords and subject terms: “male infertility”, “polymorphism” and “estrogen receptors” or “ERs”. The synonyms of polymorphism  were also used as the keywords in the search. The studies were excluded if they were not English language papers or human subject. References in retrieved articles were screened in which case reports, meta-analyses and review articles were excluded. In addition, studies were identified by a manual search of the references lists of reviews and retrieved studies. All the studies were included if they met the following criteria: (I) about the rs2234693, rs9340799, rs1256049, and rs4986938 polymorphisms and male infertility, (II) from a case control study, (III) genotype frequencies could be derived. The reasons for exclusion of articles were listed in the Additional fileTwo authors (Tian-Fu Li and Qiu-Yue Wu) extracted all data independently that met the inclusion criteria and reached the consensus for any controversy. The main characteristics of the enrolled studies were listed in the Table P <0.05.A fixed-effect model using the Mantel–Haenszel method and a random-effects model using the DerSimonian and Laird method were used to pool the results[The risk of male infertility associated with the four polymorphisms of the ERs gene was estimated for each study by odds ratio (OR), together with its 95% confidence interval (CI), respectively. The four polymorphisms were evaluated for the associations with male infertility susceptibility based on four genetic models. To contrast, the wild-type homozygote (WW), we first estimated the risk of the rare allele homozygote (RR) and heterozygous (WR) genotypes on infertility, then evaluated the risk of infertility under a dominant model (RR + WR vs. WW). In addition, recessive model associations were also estimated (RR vs. WR + WW). Moreover, stratified analyses were also performed by ethnicity (Asian and Caucasian). The statistical significance of the pooled OR was determined with the Z-test and a P-value of <0.05 was considered significant. Heterogeneity across the studies was evaluated by Chi-square test based on Q test and was  results. In addi results. Hardy–W results. All sta6/mL), azoospermia and oligoasthenoteratozoospermic (OAT). The studies’ exclusion criteria and inclusion criteria were listed in the Table A total of 10 eligible case control studies with the publication dates ranged from 2002 to 2013 met the prespecified inclusion criteria  in the Asian controls, which was higher than that in Caucasian controls 44.82% (95% CI: 41.48-48.16) as shown in Figure Pheterogeneity = 0.670), CT vs. TT , CC + CT vs. TT  and C alleles vs. T alleles  in the Asian population, as summarized in Table For rs2234693 polymorphism, significant differences were observed for the comparison of CC vs. TT, CT vs. TT and CC + CT vs. TT. Sub-group analysis by the ethnicity revealed a significant association in the comparison of CC vs. TT  in the Caucasian population, also for the AA vs. GA + GG and A alleles vs. G alleles, a significant association was observed in Asian population  as summarized in Table For rs9340799 polymorphism, increased risks were observed for the comparison of AA vs. GG and AA vs. GA + GG. Sub-group analysis by ethnicity revealed increased risks . All data were concluded in the Table Pheterogeneity = 0.818) in Asian population. For the rs4986938, there was no significant association observed in all comparisons (data were not shown).For rs1256049 polymorphism, significant differences were observed for the comparison of GA vs. GG, AA + GA vs. GG and AA vs. GA + GG. For the comparison of the GA vs. GG, AA + GA vs. GG, increased risks present in Asian and Caucasian population, respectively  . Sensitivity analysis revealed that the two independent studies[Pheterogeneity = 0.320, I2 = 14.7%). Although the genotype distributions in four studies did not follow Hardy–Weinberg equilibrium, the corresponding pooled ORs were not materially altered by excluding the studies.Among the four polymorphisms, a significant heterogeneity was apparent among heterozygote comparison for the rs1256049 . To adjust for this bias, a trim-and-fill method developed by Duval and Tweedie[Begg’s funnel plot and Egger’s test were performed to assess the publication bias of the currently available literature. For the rs2234693, rs9340799 and rs4986938, the shape of the funnel plots did not reveal any evidence of obvious asymmetry in all comparison models. Then, the Egger’s test was used to provide statistical evidence for funnel plot symmetry. The results also did not show any evidence of publication bias. However, for the rs1256049, as shown in the Figure  Tweedie was implThe present meta-analysis, including 1568 cases and 1602 controls from 10 case control studies, explored the association between the ERs polymorphisms and male infertility. The results indicated that rs2234693C allele was associated with decreased risk of the male infertility, particularly in the Asian population. In contrast, rs9340799AA genotype was observed as a risk factor for infertility in both Asian and Caucasian population, and rs1256049GA genotype was associated with an increased risk for developing male sterility. However, the rs4986938 polymorphism was not associated with male infertility. In addition, we tried to find the data in the available database, such as PUBMED, NationaEstradiol has been reported as a survival factor for germ cells, involviThe mechanisms behind altered ERs function in subjects with polymorphisms remain unclear. The polymorphism rs1256049 located at the splice acceptor site just prior to exon 8 in ERβ and may The precise role of estrogen receptors in male fertility status is understood. Some findings suggest that specific polymorphisms of the ERα, and ERβ genes which confer a lower sex hormone binding globulin (SHBG) and thus a stronger unbound estrogen effect, may adversely affect human spermatogenesis, 49. SHBIdentifying the source of heterogeneity is one of the most important goals of the meta-analysis. Thus, we stratified the studies only according to ethnicity . Stratified analysis by ethnicity revealed that there was no difference between the European population and Asian population, suggesting that different ethnicities and environmental exposures may have no influence on the susceptibility of male infertility, and more studies should be accumulated to reveal the difference. In addition, for the rs1256049, sensitivity analysis revealed that the three independent studies, 22, 23 There are still some limitations in this meta-analysis. Firstly, there were only ten literatures enrolled in this meta-analysis, the sample size was not big enough to have substantial power exploring the real association. Secondly, the detailed information  could not be traced, so that our unadjusted estimates should be confirmed by further studies. In addition, an individual with a clinical disorder was not a result of the single gene that is disrupted, but that the genetic disruption was embedded within the context of that individual's entire genome and environment exposure. In factIn summary, this meta-analysis suggested that the rs2234693C allele was the protective factor for male infertility, the rs9340799AA genotype was associated with an increased risk for infertility, and the rs1256049GA genotype was also the negative factor.Additional file 1: Text S1: The reasons for exclusion of the articles which were shown in Figure


TM6SF2 gene was recently associated with nonalcoholic fatty liver disease (NAFLD). We explored the association between E167K and plasma concentrations of alanine (ALT) and aspartate (AST) aminotransferases through a meta-analysis. We also estimated the strength of the effect across diverse liver phenotypes, including NAFLD and chronic viral hepatitis; fourteen studies were included. We found that ALT  and AST  levels were significantly associated with rs58542926 in NAFLD. By contrast, rs58542926 was not associated with either ALT  or AST  levels in four studies on chronic hepatitis. In conclusion, the results of the pooled estimates in patients with NAFLD showed that carriers of the T allele (EK + KK), when compared with homozygous subjects for the C allele (EE genotype) have increased levels of aminotransferases; however, this increase represents –2.5 (9.8%) and 1.2 (5%) IU/L of ALT and AST respectively, which is fairly small compared with the large effect of PNPLA3- rs738409-G allele that is associated with a –28% increase in serum ALT.A nonsynonymous E167K (rs58542926 C/T) variant in  Plasma concentrations of alanine (ALT) and aspartate (AST) aminotransferases have been classically regarded as markers of liver injury, including a wide range of etiologies from viral hepatitis to nonalcoholic fatty liver (NAFLD)The large body of evidence derived from epidemiological studies on the prevalence of obesity, type 2 diabetes (T2D), and cardiovascular disease (CVD) has uncovered an unexpected but biologically plausible association between aminotransferases and all of the individual components of the metabolic syndrome (MetS), as recently reviewedPNPLA3) p value for the association of ALT 1.2 × 10−45)Like all circulating substances, the plasma levels of aminotransferases are highly variable and are affected by a myriad of factors including genetic predisposition5TM6SF2 (Transmembrane 6 Superfamily Member 2), was also associated with modest p values for the association with ALT but not AST in the Dallas Heart Study and The Dallas Biobankp = 7.6 × 10−14) in the Copenhagen StudyA recent exome-wide association study of liver fat content showed that rs58542926 (E167K), a nonsynonymous variant located in Furthermore, conflicting and non-replicated results found in some1516TM6SF2 locus- and aminotransferases. Hence, whether the rs58542926 has any effect on the circulating levels of ALT or AST is still unknown. As the literature shows conflicting and inconclusive results, our primary purpose was to explore the putative association between the rs58542926 variant and plasma levels of aminotransferases by a meta-analysis of existing data. In addition, we estimated the strength of the effect of rs58542926 on both circulating ALT and AST across different populations and diverse liver phenotypes, including NAFLD and chronic viral hepatitis.Interestingly, the results from the two above-mentioned GWAS on liver enzymes8n = 11) are shown in 151617n = 4) are summarized in We evaluated fourteen studies that were identified using the search strategy described in 12161920212223242526Basic details of the included studies, including location, main clinical descriptors and sample size, and also major concerns or putative bias of the studies are summarized in Eleven studies included in the meta-analysis were hospital-based1516171819202122232412252021Genotyping for rs58542926 was carried out using a Taqman assay151718192122232426Concordance with Hardy-Weinberg equilibrium (HWE) was observed in all the studies as stated by the authors; specific assessment of departure from HWE was further performed according to the genotype frequencies reported in any paper whenever this information was not disclosed.Associations for the plasma levels of ALT were extracted from eleven studies121619202122232425261216192021232426p = 3.2 ×10−6) (p = 0.44) in a sample of 94,414 individuals of both genders. The analysis revealed a significant heterogeneity  that by the sensitivity analysis it was primarily attributed to the studies including pediatric population2021p = 0.45, I2: 0).We found that the plasma concentration of ALT was significantly associated with rs58542926 variant (random model 2 ×10−6)  without p = 0.00079, n = 93,809) (p = 0.33) but with evidence of heterogeneity  that was also successfully solved when the studies that included pediatric population20p = 0.12, I2: 36.9); the results on AST levels stratified by age are shown in Likewise, circulating AST levels were significantly associated with rs58542926 variant   without The analysis included a total of 4187 subjects from which the values of ALT according to E167K genotypes could be extracted15161715p = 0.245) models  or publication bias (p = 0.22). Neither was the rs58542926 variant associated with plasma AST levels (fixed or random model p = 0.172) in a smaller sample that included 2678 individuals without evidence of heterogeneity  or publication bias (p = 0.296)  .TM6SF2, which is involved in the regulation of lipid metabolic process, on the concentration of aminotransferases in the circulating compartment. Interestingly, by means of a comprehensive and free from bias meta-analysis of the published evidence, we found that the rs58542926 variant exerts a moderate but statistically significant effect on the circulating levels of both ALT and AST in patients with NAFLD, but not in chronic viral hepatitis.We explored the influence of rs58542926, a missense variant of The results of this meta-analysis show no evidence of publication bias; assessment of completeness of information, validity of individual studies, and analytic value of the test investigated suggest that the overall quality and methodology of studies was high. Nevertheless, some potential limitations deserve to be discussed. First, the presence of heterogeneity may potentially restrict the interpretation of the pooled risk estimates, particularly concerning the association of the variant with ALT and AST in NAFLD. However, the random effect model that does not depend on heterogeneity yielded a significant result pertaining to the association with both enzymes. More importantly, after performing a sensitivity analysis in the complete dataset, we observed that the heterogeneity was explained by three studies that enrolled pediatric population320flare-ups during the course of chronic hepatitis, could all affect the overall effect, although in a large dataset this noise may be compensated.Second, raw data of each study was not fully available; that explain why the results (rs58542926-effects) presented in our meta-analysis slightly differ from those in the original studies. In fact, we had to convert some values of medians and interquartile range or range into mean and SD in some of the studies. Third, we could not provide an estimation of the effect of the variant in the additive model because of the low frequency of the T-allele (167 K); thus, all the calculations were based on the dominant model of inheritance. Four, there were differences in the study-design between the two main liver phenotypes; while studies of patients with NAFLD were either population or hospital-based studies that included either control subjects, cases and controls, or cases only; studies of patients with chronic hepatitis were exclusively disease-centric studies that included only cases. Five, we were not able to assess the effect of the variant on the plasma levels of aminotransferases according to sex, because the studies did not disclose data of ALT or AST values separately in men and women. Finally, a limitation that is an intrinsic defect of the cross-sectional design of all the included studies, and which cannot be solved by any specific analysis, is that the values of liver enzymes incorporated into our meta-analysis correspond to the circulating measurement of aminotransferases at only one point in time. Hence, biological oscillations of transaminases, even the ones experimented during the day, after a meal, or exercisePNPLA3 rs738409-G allele that is associated with a 28% increase in serum ALT levelsTM6SF2 locus a direct role in the modulation of circulating levels of transaminases as surrogate indicators of liver damage and/or inflammation. This assumption is partially supported by the divergent results of the association of rs58542926 and aminotransferases in NAFLD vs. chronic hepatitis. Notably, compared with NAFLD, though in a smaller but adequate statistical power- sample encompassing 4187 patients with chronic viral hepatitis, the rs58542926 had no effect on the circulating levels of ALT and AST.The significant association of rs58542926 and circulating levels of aminotransferases in patients with NAFLD was observed after collecting data from a large dataset consisting of 94,414 individuals for ALT and 93,809 individuals for AST. More specifically, the results of the pooled estimates showed that carriers of the minor T allele  compared with homozygous subjects for the ancestral C allele (EE genotype) have higher levels of liver enzymes. However, it is difficult to attribute a meaningful clinical value to this finding that represents approximately an increase of ~2.5 (9.8%) IU/L of ALT and 1.2 (5%) IU/L of AST, or at least, this finding lacks of a satisfactory clinical interpretation at the population level. For instance, as mentioned earlier, compared with the large effect of the What could be the reasons for this apparent discrepancy between the effect of the E167K variant in NAFLD and chronic viral hepatitis?p = 0.0001) between the mean ALT value in the studies of patients with NAFLD (25.3 ± 19.5 IU/L) compared that of the studies of patients with chronic hepatitis (72.49 ± 167.82 IU/L). Similarly, the plasma levels of AST significantly (p = 0.0001) differ between the studies of patients with NAFLD (23.96 ± 15.3 IU/L) versus studies of patients with chronic hepatitis (51.98 ± 124.91 IU/L). These differences probably reflect a distinctive clinical meaning of the elevated levels of these enzymes in each of the two phenotypes. For instance, in patients with chronic viral hepatitis, the elevation of aminotransferases in the order of ~ two times or more the upper normal limit most likely represents histological changes associated with hepatocellular injury. On the contrary, circulating ALT or AST values in NAFLD, which are hardly ever in the range of chronic hepatitis and can be even in the “normal” range, do not necessarily represent histological severity but rather metabolic perturbations430433In patients with chronic hepatitis, rather than liver steatosis, liver damage, inflammation, and necrosis are part of the main histological picture; then, plasma levels of aminotransferases probably reflect liver injury more accurately than metabolic perturbations. An interesting aspect to highlight that reinforces the previous assumption is the significant difference  higher hepatic fat content when compared with carriers of the EE genotypehttp://ncbi.nlm.nih.gov/entrez/query), Google Scholar and EMBASE and the Science Citation Index databases were performed using the search terms “TM6SF2” in all fields, and “rs58542926, gene or variants or polymorphism or alleles”. In addition, citations in retrieved articles as well as articles disclosed by the PubMed “related articles” link were further evaluated for inclusiveness. All the published evidence until March 2016 without any country restriction was included in the initial assessment; 48 studies were selected for the analysis.Electronic searches of Pubmed at the National Library of Medicine (The search and the assessment of the eligibility criteria was conducted by the authors (SS and CJP), who independently performed this task; there were no discrepancies in this process and the inter-observer variability by calculating the kappa statistic was 1.0. A detailed description of the process of data collection, including identification, screening, eligibility and selection of the studies can be found in the http://www.cdc.gov/genomics/hugenet/participate.htm), which are specific for the assessment of the strength of evidence for gene-phenotype associations.This meta-analysis followed the appropriate methods for conducting the meta-analysis of genetic association studies, as stipulated in The Human Genome Epidemiology Network (HuGENet) guidelines  allele (EE genotype) vs. the carriers of the T allele (Lys167), specifically the heterozygous EK+ homozygous KK genotypes, as explained earlierFrom each study included in our meta-analysis, we retrieved relevant information on demographics, such as age, sex, and ethnicity, and circulating levels of ALT and AST expressed in international units (IU/L) and measured by any standard analytical method ; standard error and interquartile rank were converted to SD. Data on the explored phenotypes was extracted for the EE and EK+KK genotypes, and the analyses were based on comparing the genotype groups without any further adjustment for confounding factors.I2 statistic, which is a transformation of Q that estimates the proportion of the variation in effect sizes that is due to heterogeneity between the studies. In such cases, an I2 value of 0% indicated no observed heterogeneity, while greater values corresponded to increasing heterogeneity. In the case of heterogeneity, we identified study characteristics that stratified the studies into subsets with homogeneous effects. We considered the possible sources of heterogeneity and stratified the studies by age and ethnicity, and also repeated the analysis separately for each group. If the association became homogeneous after stratification or after excluding the outlier studies, we recalculated the overall effect and 95% CI, and no further action was taken. Although the studies excluded in this way cannot be considered outliers, excluding studies that contribute the most to heterogeneity is an unbiased way of achieving the homogeneity required for a stringent meta-analysisp value of ≤0.05 was considered statistically significant. All calculations were performed using the Comprehensive Meta-Analysis computer program .While the main outcome (ALT and AST levels) was measured in IU/L across all studies, the expression of data was not always uniform, and potential variability across different laboratories could also be expected. Then, we attempted to homogenize the results through the use of standardized mean difference (D), which is the difference in means pertaining to cases and controls divided by the common within-group SD. The difference in means was used to estimate the magnitude of the absolute effect in IU/L. The statistical protocol applied to this study was already published in our previous work on the meta-analysis of the role of rs58542926 on lipid traits and NAFLDHow to cite this article: Sookoian, S. and Pirola, C. J. Meta-analysis of the influence of TM6SF2 E167K variant on Plasma Concentration of Aminotransferases across different Populations and Diverse Liver Phenotypes. Sci. Rep.6, 27718; doi: 10.1038/srep27718 (2016).