INK4ap16 is an important tumor suppressor gene (TSG) and aberrant methylation of promoter is known to be a major inactivation mechanism of the tumor suppressor and tumor-related genes. Aberrant TSG methylation was considered an important epigenetic silencing mechanism in the progression of head and neck squamous cell carcinoma (HNSCC). However, some studies have reported differences in the methylation frequencies of 16INK4aP promoter between cancer and the corresponding control group. Therefore, we conducted a meta-analysis to better identify the association.The INK4ap16 promoter methylation and HNSCC. Odds ratio (ORs) and 95% confidence intervals (95%CI) were calculated to evaluate the strength of association between INK4ap16 promoter methylation and HNSCC.PubMed, Ovid, ISI Web of Science, and EMBASE were searched to identify eligible studies to evaluate the association of INK4ap16 promoter methylation in the cancer group were significantly higher than those in the control group . The pooled odds ratio was 3.37 (95%CI = 2.32–4.90) in the cancer group versus the corresponding control group under the random-effects model.A total of twenty-one studies with 1155 cases and 1017 controls were included in the meta-analysis. The frequencies of INK4ap16 promoter was found to be significantly associated with HNSCC.This meta-analysis of 21 published studies identified that aberrant methylation of  Head and neck squamous cell carcinoma (HNSCC) occurs in the oral cavity, oropharynx, hypopharynx and larynx. HNSCC is the sixth most common cancer worldwide and the fifth leading cause of cancer death. Over 500,000 HNSCC cases occur each year . Howeverp16INK4a gene plays a key role in cell cycle regulation and is located on chromosome 9p21 which consisting of three exons and two introns, spanning approximately 8.5 kb [p16INK4a promoter may play an important role in the development of HNSCC.The pathogenesis of HNSCC is a multistep and multifactorial complex mechanism containing a variety of genetic and epigenetic abnormalities, signal transduction, apoptosis, angiogenesis, and cell cycle regulation . Epigeney 8.5 kb . P16 prep16INK4a promoter between cancer and non-cancerous. However, the results are inconsistent. Therefore, the aims of this meta-analysis are to consolidate the available data and to better identify the association between p16INK4a promoter methylation and HNSCC.Some studies have reported differences in the methylation frequencies of Relevant studies were identified from the online electronic databases  using the search terms: (squamous cell carcinoma or cancer) and  AND (p16 methylation). The search was limited to English language paper. The search results were updated until August 20, 2014.p16INK4a promoter methylation frequency and HNSCC, (2) case-control study or providing the case and the control cases, (3) providing the p16INK4a promoter methylation frequency in case and control groups, (4) specimens of HNSCC were surgically respected primary tumor sample. Firstly, the titles and abstracts of initial searching articles were evaluated for whether it met the inclusion criteria. Then all potentially relevant articles were evaluated on full-text paper. If the results of a study were published more than once, only the most complete and up-to-date information were included in the meta-analysis. The study selection process was shown in A study included in the meta-analysis had to meet the following criteria: (1) studies with evaluating the association between p16INK4a methylation in the case and control groups, the number of the case and control groups. All the detailed information of the included studies was checked by three reviewers  as described in the Cochrane Handbook for systematic reviews.Three reviewers  independently reviewed the included articles. The following information was extracted from the included studies: first author’s name, year of publication, study population, sample size, sample type in case and control group, the state of the individual of the control group, the number of people with p16INK4a promoter methylation and HNSCC. The x2-based Cochran Q statistic test and I2 statistics were used to test the between-study heterogeneity [P<0.05 for the Q statistic or I2 ≥50%, the heterogeneity was considered significant and a random-effects model was used to calculate the pooled ORs. Otherwise, a fixed-effects model was applied. The meta-regression (method = restricted maximum-likelihood estimator) was employed to explore the source of heterogeneity. And the subgroup analysis was performed for further evaluation of the source of heterogeneity. The contribution of each study to the final results of the meta-analysis was evaluated according to the sensitivity analysis. Begg’s funnel plot [P values are two sides with a significant level at 0.05. When the individual studies have cells with zero counts, the default is to add 0.5 to all zero counts in the Meta package. All statistical analyses were performed with the Meta package  in R .The pooled odds ratios (ORs) and their 95% confidence intervals (CIs) were used to assess the strength of the association between nel plot  and Eggenel plot  were usenel plot  was alsop16INK4a promoter methylation frequency and HNSCC. A total of 21 studies with 1155 cases and 1017 controls were included in the meta-analysis. 397 studies were initially identified by searching the electronic databases. 63 potentially relevant studies were retrieved for further evaluation after removing 82 duplicated articles, 59 reviews or meeting reviews, 9 cell lines and animal studies, 175 irrelevant articles, and 9 papers that did not have full text versions. 32 studies without a control group, 8 studies without p16INK4a promoter methylation data, and 2 studies without tissue in the case group were excluded from full-text review. Finally, 21 studies were included in our meta-analysis. Ten studies were of Asian subjects and eleven studies were of Caucasian subjects. The control group included HNSCC patients, benign disease patients, and healthy people. And the sample type of control group included tissue, serum, blood, and saliva. The methylation detection method of p16INK4a in HNSCC and control included 16 studies used methylation-specific polymerase chain reaction (MSP), three studies used real-time quantitative MSP (QMSP), one study used PCR-based methylation-sensitive restriction analysis (MSRA), and one study used pyrosequencing (Pyro). Six studies [The comprehensive search and selection procedures of literature were displayed in  studies –18 were  studies –33 were p16INK4a promoter methylation with HNSCC risk in I2 = 60.1%, Q = 50.07, P<0.001). In the overall meta-analysis, p16INK4a promoter methylation frequency was significantly associated with HNSCC  (The combining result showed the association of 32–4.90) .p16INK4a promoter methylation and HNSCC were reliable and stable.A sensitivity analysis was performed by omitting a single study and calculating the pooled OR for the remaining studies under the random-effect model to assess the effects of each individual study on the pooled OR. The results of sensitivity analysis were summarized in the I2 = 60.1%, Q = 50.07, P<0.001), the meta-regression was employed to explore potential sources of heterogeneity. We conducted a multiple regression model with six variables . No source of significant heterogeneity was found . Subgroup analysis of the control sample type showed that the frequencies of p16INK4a promoter methylation in the Non-tissue group  were higher than those in the tissue group (control sample type: tissue) under the random-effects model . For the methylation detection method of p16INK4a promoter, Ghosh [The OR for was 4.76 (95%CI = 3.00–7.54) in Asians under the random-effects model, and 2.25 (95%CI = 1.61–3.13) in Caucasians under the fixed-effects model. With the case sample size, the OR for was 3.33 (95%CI = 1.68–6.60) in the >60 case group under the random-effects model, and 3.59 (95%CI = 2.65–4.87) in the ≤60 case group under the fixed-effects model. In the subgroup analysis of the control type, the frequencies of r, Ghosh  used PCRP = 0.25). The fail-safe number  also did not display statistical evidence for publication bias.Begg’s funnel plot, Egger’s test, and the fail-safe number were performed to assess the publication bias of the literature. The shape of the Begg’s funnel plot in TSG) promoter can be found in various cancers and induce the silencing of tumor suppressor genes in the process of carcinogenesis. The p16INK4a is an important tumor suppressor and DNA repair genes with a major negative regulator of critical tumor pathway [p16INK4a induced by aberrant hypermethylation plays a role in the process of carcinogenesis in various cancers  [Hypermethylation of the tumor suppressor gene ( pathway  and the  pathway . Inactiv cancer) –39.p16INK4a promoter methylation and HNSCC, which included 21 studies with 1155 cases and 1017 controls. The frequency of p16INK4a promoter methylation in tumor was 43.98% and 21.14% in control. The pooled odds ratio under a random-effect model was 3.37 (95%CI = 2.32–4.90) in the cancer cases versus the controls. The result identified that methylation of p16INK4a promoter had 3.37 fold increased risks of HNSCC compared with the control group.The current meta-analysis assessed the association between p16INK4a promoter methylation and HNSCC in Asians was stronger than that in Caucasians. Similarly, Kaur’s study observed that the frequency of p16INK4a promoter methylation (47.8%) in Indian cohort was higher than that in North-American cohort (37.5%) [In subgroup analysis, the OR for was 4.76 (95%CI = 3.00–7.54) in Asians under the random-effects model, and 2.25 (95%CI = 1.61–3.13) in Caucasians under the fixed-effects model. The association between  (37.5%) . However (37.5%) . The dif (37.5%) . Pyro is (37.5%) . In addip16INK4a promoter methylation was 18% (20/113) [p16INK4a promoter methylation between HPV16 positive cancer and HPV16 negative cancer [Some studies found that reduced p16 expression in 55–89% of oral and oropharyngeal SCC –46. The (20/113) . Human p(20/113) –54, howee cancer . Then, fI2 = 60.1%, Q = 50.07, P<0.001) in the meta-analysis. According to meta-regression and subgroup analysis, no sources of heterogeneity were founded. Therefore, some source of heterogeneity may exist among studies. Secondly, we do not study the association between p16INK4a promoter methylation and disease characteristics  of HNSCC. According to study the association between p16INK4a promoter methylation and disease characteristics, the special role of p16INK4a promoter methylation in the carcinogenic process of HNSCC may be found.There were some limitations in the meta-analysis. Firstly, a significant heterogeneity across the studies was found Click here for additional data file.S2 Checklist(DOCX)Click here for additional data file.


Adiponectin is an insulin-sensitizing hormone produced by adipocytes. It has been suggested to be involved in endometrial tumorigenesis. Published data have shown inconsistent results for the association between circulating adiponectin levels and endometrial cancer. In this study, we conducted a meta-analysis to evaluate the predictive value of circulating adiponectin levels on the development of endometrial cancer.PubMed, Embase, ISI web of knowledge, and Cochrane databases were searched for all eligible studies, and the summary relative risk (SRR) was calculated. Additionally, we performed dose-response analysis with eight eligible studies.A total of 1,955 cases and 3,458 controls from 12 studies were included. The SRR for the ‘highest’ vs ‘lowest’ adiponectin levels indicated high adiponectin level reduced the risk of endometrial cancer . Results from the subgroup analyses were consistent with the overall analysis. The SRR for each 1 µg/ml increase of adiponectin indicated a 3% reduction in endometrial cancer risk (95% CI: 2%–4%), and a 14% reduction for each increase of 5 µg/ml (95% CI: 9%–19%). No evidence of publication bias was found.This meta-analysis demonstrates that low level of circulating adiponectin is a risk factor for endometrial cancer. A low p-value of Cochran’s Q-statistic (<0.05) indicates significant heterogeneity among studies. To investigate the effect of potential confounders, we conducted subgroup analyses by considering all the possible factors, such as study population, menopausal status of individuals, and adjustment factors for age, BMI, hypertension and diabetes. To evaluate the stability of the pooled estimates, we performed sensitivity analysis by examining changes in results after sequential omission of individual studies.We combined both odds ratios (ORs) and relative risks (RRs), for calculating the summary relative risk (SRR). The multivariate-adjusted risk estimates and corresponding confidence intervals were transformed into log relative risks for meta-analyses . Pooled I2 test  and CochI2 test  were useThe summary estimate of the dose-response effect of adiponectin levels on endometrial cancer risk was calculated using fixed-effects model. Generalized least-squares regression was utilized to estimate the linear trend in SRR, while the method of restricted cubic spline was applied for nonlinear trend estimation . PublicaA total of 12 eligible studies –21,30 inI2 = 67.5%, p = 0.001). Therefore, we conducted subgroup analysis to evaluate the influence of single factor that might have impact on the result. I2 = 53.8%) [I2 = 64.8%) [I2 = 53.9%) [I2 = 64.2%) [I2 = 58.8%) [I2 = 0.0%) [I2 = 29.1%) [I2 = 0.0%) [The SRR for the “highest” vs “lowest” category of adiponectin levels was 0.47 , which indicated a 53% reduction in endometrial cancer risk  11–21,3,321,30. = 53.8%) ,16,18–20= 64.8%) ,17,18,21= 53.9%) ,16,19,20= 64.2%) –17,19,20= 58.8%) –15,17,20 = 0.0%) ,19, in s = 0.0%) ,16.We evaluated the dose-response effect of adiponectin levels on endometrial cancer risk. An inverse linear dose-response correlation was observed from pooled 8 studies (p < 0.001), with a 3% (95% CI: 2%-4%) reduction of endometrial cancer risk for each 1 g/ml increase of adiponectin, and a 14% (95% CI: 9%-19%) reduction for each 5 g/ml increase of adiponectin. Additionally, a statistically nonlinear correlation between adiponectin level and endometrial risk was also observed (p < 0.001). No evidence of heterogeneity was observed when we calculated either the linearity (p = 0.10) or the nonlinearity (p = 0.25).We conducted a sensitivity analysis in which individual studies were sequentially omitted and the rest analyzed. The results indicated none of the single studies significantly affected the SRRs. We did not detect any publication bias by the Egger’s (p = 0.497) or Begg’s test (p = 0.533).et al., but it was not significant which might due to the limited number of subjects analyzed [Our meta-analysis suggests an inverse correlation between circulating adiponectin level and endometrial cancer, with a 53% reduction in risk for higher levels. The result was consistently significant in all sensitivity analyses and in all populations studied. However, no significant associations were observed in premenopausal women, and in studies not adjusted for age. The dose-response analysis shows that each 1 g/ml increase of adiponectin corresponds to 3% reduction in risk of endometrial cancer. The overall dose-response association was estimated by Petridou analyzed . MoreoveIn our stratified analysis by menopausal status, we only found a statistically significant association between adiponectin level and endometrial cancer risk in postmenopausal women. Indeed, three studies including only 297 subjects reported nonsignificant results among premenopausal women ,13,19. Tet al. also observed that high-glycemic load diets, which cause high levels of blood glucose and insulin, are directly related to endometrial cancer risk [in vitro [Many studies have addressed on the mechanisms for the role of adiponectin on reducing endometrial cancer risk. One major mechanism is that adiponectin can decrease blood insulin levels ,33–35, acer risk . Inflammcer risk . Adiponecer risk ,39–44. Fcer risk . Moreovein vitro . Finallyin vitro . It has in vitro . Therefoin vitro –17, whicIn addition, adiponectin circulates in plasma in three forms of low-molecular-weight (LMW) oligomeric complex of a trimer, a middle-molecular-weight (MMW) complex of two trimers and a high-molecular-weight (HMW) complex of up to six trimers . ExperimThe current meta-analysis supports the direct link between lower level of circulating adiponectin and increased endometrial cancer risk. By combining information from all relevant studies, our meta-analysis overcomes the limitations of time period and the specific subject population of study, increases the sample size and thus the power to study effects of interest. Although endometrial cancer is a frequent gynecologic cancer, it is a relatively low-incidence disease compared to other common cancers such as breast cancer, lung cancer and colorectal cancer, odds ratio (OR) can be used as a good estimate of relative risk. Therefore it is appropriate to combine both ORs and RRs for calculating a summary relative risk. However, this study still has several limitations. First, selection bias may exist, because parts of control subjects are hospital-based, who might have benign disease and different risks for developing endometrial cancer. Second, adjustments for the clinicopathological parameters, which may result in measurement error, together with other inter-study variances enhance the between-study heterogeneity. Third, some of the findings in subgroups are limited by small sample sizes. Further large prospective cohort studies should be carried out to investigate the optimal level of adiponectin with a protective effect on endometrial cancer. Although various pharmacological and non-pharmacological interventions have been adopted to influence adiponectin levels, the clinical relevance of the elevated adiponectin level to the reduction of the endometrial cancer risk remains to be determined .et al. which showed that RR is in the opposite direction as we hypothesized Click here for additional data file.


Mutations of glucocerebrosidase (GBA) confer susceptibility to Parkinson's disease in several ethnical populations, with a high incidence especially in the Ashkenazi Jewish population. Although there are several studies that have investigated a similar association in a Chinese population, small sample sizes and few positive outcomes have made it difficult to obtain conclusive results from these individual studies. Therefore, the present study used a meta-analysis approach, pooling the appropriate data from published studies to investigate the association of GBA mutations and Parkinson's disease in a Chinese population. Nine studies containing 6536 Chinese subjects  and examining the GBA mutations of L444P, N370S and several other mutations were included. Review Manager 5.2 software was applied to analyze the pooled odds ratios (ORs) and 95% confidence intervals (CIs). The results showed a significant association of Parkinson's disease risk with overall GBA mutations , and with the subgroup of L444P mutation . No such association was observed for the subgroup with N370S mutation or other mutations, in part because of the small sample size or rare events. Thus, for the rare occurrence of GBA mutations, studies with larger sample size are necessary to minimize the sampling error and to obtain convincing results. Parkinson disease (PD), which ranks second only to Alzheimer's disease among neurodegenerative disorders, is characterized pathologically by the degeneration of dopaminergic neurons in nigrostriatal system and clinically by insidiously progressive movement impairments, such as rigidity, bradykinesia, impaired balance and tremor at rest Mutations of glucocerebrosidase (GBA) gene which located at chromosome 1q21 and encoding the enzyme glucocerebrosidase This study was conducted according to the Preferred Reporting Items for Systematic Reviews and Meta-analyses (PRISMA) criteria. Electronic databases, such as PubMed, Embase, Cochrane Library and PDgene, were systemically searched for eligible records, with no lower date limit but an upper limit of February 20, 2014. A combination of key words was used in the search strategy, such as “glucocerebrosidase” or “GBA”, “polymorphism” or “variant” or “mutation”, “Parkinson's disease” or “PD” and “Chinese”. The references within relevant reviews were manually examined to avoid missing eligible studies. The language was not restricted in the search.Studies included in the meta-analysis met the following criteria: (1) investigated the association between GBA mutations and PD in a Chinese population; (2) contained full text within the article; (3) contained sufficient data to calculate the odds ratio (OR), confidence interval (CI) and p-value; (4) specified genotyping methods or provided appropriate references; and 5) The diagnosis of PD should be based on established clinical evidence , or meet UK PDS Brain Bank Clinical Diagnostic Criteria Information was extracted from all included publications by two independent reviewers using a predesigned data extraction form. Disagreements were resolved by consensus. If the two reviewers could not reach a consensus, another reviewer was consulted. The following information was collected from each article: first author, publication date, study region, number of cases and controls, and the number of GBA mutations in PD cases and in controls.2) test was used to assess the heterogeneity across studies, and 2I statistics were calculated to quantify the proportion of total variation due to the heterogeneity. The ORs were pooled with a fixed effects model using the Mantel–Haenszel approach when no heterogeneity was observed among the studies Because the frequency of GBA mutations is low, the Hardy–Weinberg equilibrium NciI, R120W, D409H, L174P, Q497R and V460M, were also investigated in several studies, but rarely showed positive outcomes.2 = 9.37, I2 = 15%, p = 0.31); thus a fixed effect model was used to calculate the pooled ORs and 95% CIs. The results of the statistical analysis revealed a significant difference in the GBA mutation-related susceptibility . Intriguingly, mutations of GBA gene have been expected to confer a susceptibility to Parkinson's disease, which shows a high rate of comorbidity with GD NciI, R120W, D409H, L174P, Q497R and V460M, were also investigated in several studies. However, almost no positive mutation was detected in the majority of these studies, which likely due to their small sample sizes, which brought low statistical powers and less reliable results.Several studies have examined the relationship between GBA mutations and PD risk in Chinese population. To draw a comprehensive understanding, we pooled the data, performed a meta-analysis and reported the distribution of the studied GBA mutations in Chinese population. The meta-analysis included 6536 subjects, with 3438 cases and 3098 healthy controls; however, only 125 mutations in GBA cases  and 15 in controls  were detected. These results suggested a low rate of GBA mutation in Chinese population . The mutation rate of GBA in patients from the Chinese population (3.64%) was also lower than that in patients from Ashkenazi Jews cohort(19.61%) and non-Ashkenazi Jews  Although primary results of the present retrospective analysis are compelling, this study has a few limitations. First, data included in the meta-analysis were obtained from published articles. We could not track unpublished articles, influencing the comprehensiveness of the data. Second, a language bias might have occurred as some articles published in languages other than English that might have been missed out. Third, the sample size in some of the studies included in our analysis was too small to enhance the power of the meta-analysis. Finally, within-study confounding factors, such as genetic background, might have existed. The Chinese comprise multinational populations with probable genetic differences. In addition, the Chinese subjects recruited in Singapore may have introduced confounding genetic factors.Taken together, our meta-analysis showed a significant association between GBA mutations and the susceptibility to Parkinson's disease in a Chinese population. Specifically, we demonstrated that the L444P mutation was a positive risk factor in PD. Another common mutation, N370S, failed to produce a significant association with PD, likely due, at least in part, to the small sample size. Thus, additional studies examining the association of GBA mutations, especially N370S and other rare mutations, with PD risk in the Chinese population with large sample sizes are needed.S1 TableStatistical power for each GBA mutation locus.(DOCX)Click here for additional data file.S1 ChecklistPRISM 2009 checklist.(DOC)Click here for additional data file.S1 FilePRISM 2009 flow diagram.(DOC)Click here for additional data file.


The relative risks (RRs) for non-smokers, former smokers and current smokers were 0.89 (95% CI: 0.63–1.27), 0.74 (95% CI: 0.62–0.89) and 0.86 (95% CI: 0.79–0.93), respectively. The results remained stable in subgroup analyses by other confounders and sensitivity analysis. Conclusions: The results of our meta-analysis suggest that a healthy dietary pattern is associated with a lower lung cancer risk, and they provide more beneficial evidence for changing the diet pattern in the general population.Background: Diet and nutrients play an important role in cancer development and progress; a healthy dietary pattern has been found to be associated with several types of cancer. However, the association between a healthy eating pattern and lung cancer risk is still unclear. Objective: Therefore, we conducted a systematic review with meta-analysis to evaluate whether a healthy eating pattern might reduce lung cancer risk. Methods: We identified relevant studies from the PubMed and Embase databases up to October 2015, and the relative risks were extracted and combined by the fixed-effects model when no substantial heterogeneity was observed; otherwise, the random-effects model was employed. Subgroup and publication bias analyses were also performed. Results: Finally, eight observational studies were included in the meta-analysis. The pooled relative risk of lung cancer for the highest  Lung cancer is one of most common cancer forms worldwide, and it ranks as a top cause of cancer death . CigaretIt has gradually been recognized that diet and nutrients play an important role in cancer development and progress, and many dietary components are found to be associated with cancer risk . With reet al. first reported the association with null results in the overall population [A healthy dietary pattern  is characterized by a high intake of vegetables, fruits, white meat, fish and whole-grain breads and a low intake of red meat, high fat and refined grains . Previoupulation , and thepulation ,25,26,27pulation ,29,30. FWe conducted and reported our meta-analysis in accordance with the Meta-analysis of Observational Studies in Epidemiology (MOOSE) guidelines . Relateda posteriori method), including factor analysis and cluster analysis, or an index-based approach using dietary guidelines (a priori method). In our study, healthy dietary patterns identified by two approaches were all included if they met the following criteria: (1) cohort or case-control design; (2) the outcome was lung cancer risk, including both incidence and mortality; (3) relative risk (or odds ratio or hazard ratio) with 95% confidence intervals could be obtained or calculated from data in the manuscript. When several publications were reported on the same population, the most recent and informative one was preferred in our meta-analysis.Our interests were focused on the studies investigating the relationship between multiple food groups and lung cancer risk. There is still no accurate definition of a healthy dietary pattern, which is a relatively recent concept, in contrast with “western dietary  pattern” or “drinker dietary pattern”. The dietary patterns can be classified by a data-driven approach (Two investigators (Yanlai Sun and Zhenxiang Li) independently reviewed all the studies and extracted the following information from the identified articles, including the first author’s name, publication year, duration, location, design, dietary assessment, dietary pattern identification method, RR with 95% CIs and adjusted variables. We also evaluated the quality of included studies by the Newcastle-Ottawa scale, which is widely used in assessing observational studies . BrieflyI2 statistics; for the Q statistic, if p < 0.1, significant heterogeneity was considered to exist, and in this case the random-effects model was used to pool the original RRs. Otherwise, the fixed-effects model was employed. We tried to conduct subgroup analyses stratified by confounding factors to test the stability of the results, and also performed a sensitivity analysis in which the individual study was removed to examine the influence of a specific study on the overall results. Publication bias was evaluated by the use of the funnel plot and Egger’s test, with p < 0.1 representing a significant publication bias. All statistical analyses were conducted with STATA 12.0 software . As described before, a healthy dietary pattern is defined by a combination of food groups, and can be varied a bit among different studies, which might also be labeled as “healthy” ,26,27,29et al. [Initially, 4276 records were found in the electronic databases; after a process of careful checking and review, a total of eight studies, involving 13,548 lung cancer cases and 108,748 participants, were finally included in our meta-analysis. They included four case-control studies and four cohort studies. The literature search flow is shown in et al.  only cheet al. ,30.a posteriori data-driven approaches, such as principal component factor analysis [a priori methods based on the HEI -2010 Score [Overall, the quality scores of the original studies ranged from 6 to 9, with an average of 7.38 points. Eight of nine studies were judged as high quality, and no low quality studies were found. All the studies employed food frequency questionnaires (FFQs) in dietary assessment, and five of them were validated ,25,27,28analysis ,28,29,30analysis , and the10 Score . The resI2 = 14.5%, Q = 8.18, P = 0.32); thus, the fixed-effects model was used. A summary RR of 0.81 (95% CI: 0.75–0.86) was yielded after combining all the results , and the pooled RR for studies using validated FFQs was 0.76 (95% CI: 0.65–0.88), indicating the robustness of the results , while it disappeared in never smokers ; possibly due to the small number of the sample size, the pooled RR for current smokers was 0.86 (95% CI: 0.79–0.93) . After s results .et al. [et al. [Not enough data were obtained on the histological subtypes of lung cancer; however, the studies by Gnagnarella et al.  and De S [et al.  in 2011 I2 = 0.00%, Q = 4.9, p = 0.56) in this scenario (p = 0.18), although the funnel plot seemed slightly unsymmetrical , and a broad time span from 1986 to 2010. Additionally, subgroup analyses by gender, smoking status and dietary pattern analysis method were also performed to minimize the effects of confounding factors. et al. [However, some shortcomings should be also recognized in our study. Firstly, different methods for dietary assessment and pattern analyses were employed in the original studies; particularly, the FFQs used in several studies ,29,30 weet al. , and wheSecondly, recall and selection bias were inevitable in the observational studies, especially for hospital-based case-control studies. Inadequate adjustment for potential confounders including social-economic status, physical activity, and total energy intake also influenced the reliability of the results, because healthy eating might be a sign of other aspects of healthy lifestyles, including more exercise, positive attitude, better medical service, and less frequent smoking and drinking. In addition, given the limited number of studies included and insufficient data, subgroup analyses by histological types and ethnicity were impossible to carry out. Especially there is still no data from Asia, including China, Japan or India. Notably, compared to the Western dietary pattern, the Asian food pattern generally may fall under “healthy diet”. Thus, more studies to explore the relationship between Asian diet and lung cancer risk are required.Lastly, since our literature was limited to publications in English, and studies with null results tend to not be published, this might cause incomplete articles to be included in the meta-analysis; however, no significant publication bias was found by Egger’s test, suggesting the influence was acceptable. I2 = 59.8%, p = 0.04), indicating that bad eating habits might increase lung cancer risk significantly. In addition, the Mediterranean diet pattern, which is considered a special type of healthy eating pattern, has also been found to reduce lung cancer risk in two studies [In contrast with healthy pattern, we also attempted to combine all the results on the association between Western dietary pattern and lung cancer risk in five published articles ,29,30,33 studies ,27. ThusIn summary, findings from the meta-analysis of observational studies suggest that a healthy dietary pattern is associated with lower lung cancer risk and provides more beneficial evidence for changing the diet pattern in the general population. Given the limited number studies included, more prospective studies with strict control of confounders or interventional trials, especially in Asia, are needed to confirm this association.


Background: The present study was aimed to examine the possible association between methylene tetrahydrofolate reductase (MTHFR) gene polymorphisms and childhood acute lymphoblastic leukemia (ALL) in a sample of Iranian population.Subjects and Methods: A total of 220 subjects including 100 children diagnosed with ALL and 120 healthy children participated in the case-control study. The single nucleotide polymorphisms (SNPs) of MTHFR were determined by ARMS-PCR or PCR-RFLP method.Results: Our investigation revealed that rs13306561 both TC and TC + CC genotypes decreased the risk of ALL compared to TT genotype . In addition, the rs13306561 C allele decreased the risk of ALL in comparison with T allele . MTHFR rs1801131 (A1298C) polymorphism showed that the AC heterozygous genotype decreased the risk of ALL in comparison with AA homozygous genotype . Neither the overall Chi-square comparison of cases and control subjects  nor the logistic regression analysis showed significant association between C677T polymorphism and ALL .Conclusion: The current investigation findings showed that MTHFR rs1801131 and rs13306561 polymorphisms decreased the risk of ALL in the population which has been studied. Further studies with larger sample sizes and different ethnicities are required to validate our findings. It is also involved in formation of S-adenosylmethionine (SAM), the universal donor of one carbon unit for the methylation of different substrates.6-8 Two common polymorphisms in MTHFR gene have been described: First, C to T transition at nucleotide 677 , which causes the substitution alanine with valine at codon 222 and decreases in the enzymatic activity and subsequently increases the level of homocysteine and altered distribution of folate.        9  Second, A to C transition at position 1298 , which results in the substitution of glutamate to alanine at codon position 429 and also leads to reduced MTHFR activity.        10  Pediatric cancer is currently the second most common cause of cancer death in children who live in developed countries.11-14 The present study aimed to find out the possible association among rs1801131 (A1298C), rs1801133 (C677T) and rs13306561 polymorphisms in the MTHFR gene and ALL risk in a sample of the Iranian population.Several genetic association studies have evaluated the associations between ALL and MTHFR gene polymorphisms in various populations and produced contradictory or inconclusive results.Patients 15-18 Local ethics committee of Zahedan University of Medical Sciences was approved the project and informed consent was obtained from parents of cases and controls . DNA was extracted from peripheral whole blood using salting-out method.19This case-control study has been done in children diagnosed with ALL (n=100) and control individuals (n=120), who were matched for age and sex in southeast of Iran (Zahedan). The study design and the enrolment procedure have been described previously.GenotypingGenotyping of MTHFR rs1801133  and rs13306561 gene polymorphisms was done by PCR-RFLP method. For rs1801133 variant, the forward and reverse primers were 5′-ACTCTCTCTGCCCAGTCCCTGTG-3′ and 5′-AAGAACTCAGCGAACTCAGCACT-3′, respectively.2O were added. The PCR conditions were set as follows: 95˚C for 5 min, 30 cycles of 95˚C for 30s, 64˚C for 30s and 72 ˚C for 30s and a final extension step of 72 ˚C for 10 min. Then, 10 µl of PCR product is digested with HinfI restriction enzyme. The T allele was digested and produced 222-bp and 167-bp fragments, while the C allele was undigested and produced 389-bp product.In each 0.20 ml PCR reaction tube, 1 µl of genomic DNA (~100 ng/ml), 1 µl of each primer, 10 µl of 2X Prime Taq Premix  and 7 µl ddHFor MTHFR rs13306561 polymorphism, a set of forward and reverse primer pairs was 5′-GGCCGAGCGTTCTGAGTC-3′ and 5′-CACTAATCCCGCGAAGGGTGCTCA-3′, respectively. The PCR cycling conditions were the initial denaturation step at 95 °C for 5 min, followed by 30 cycles of 95 °C for 30s, 64 °C for 30s, and 72 °C for 30s with a final extension of 72 °C for 10 min. Ten microliters of the PCR product were digested by HpyF3I restriction enzyme. The C allele was digested and produced 162-bp and 23-bp, while the T allele was undigested (185-bp).Genotyping of MTHFR rs1801131 (A1298C) polymorphism was done using allele specific polymerase chain reaction (AS-PCR). The primer sequences were as follow: generic primers, 5′-AGACCTTCCTTGCAAATACAT-3′; A allele, 5′-GAAGACTTCAAAGACACTCT-3′; C allele, and 5′-GAAGACTTCAAAGACACTCG-3′.The beta-2 microglobulin  was used as internal control.The PCR cycling conditions were 95˚C for 5 min, 33 cycles of 95˚C for 30s, 55˚C for 20s, and 72˚C for 30s, followed by a final extension step for 10 min at 72˚C. The products size were 317-bp for either of the A or C alleles and 574-bp for the internal control.Statistical analysis20 A p-value less than 0.05 were considered statistically significant.Statistical analysis was done using statistical package SPSS 20 software. Data were analyzed by independent sample t-test and χ2 test. The association between MTHFR gene polymorphisms and ALL was estimated by computing the odds ratio (OR) and 95% confidence intervals (95%CI) from logistic regression analyses. Haplotype analysis was performed using SNPStats software. The study group involved 100 ALL patients  and 120 healthy children . No significant difference was found between the groups concerning sex and age .The genotype and allelic frequencies of MTHFR gene polymorphisms in cases and controls are presented in Both rs13306561 TC and TC + CC genotype decreased the risk of ALL compared to TT genotype . On the other hand, C allele decreased the risk of ALL in comparison with T allele . We also analyzed the MTHFR polymorphisms interaction . In compHaplotype analysis results are shown in  ALL is a multifactorial disease influenced by genetic and environmental factors. In the current study, we examined the impact of MTHFR gene polymorphisms on the risk of childhood ALL in a sample of the Iranian population. Our findings indicated that rs1801131 (A1298C) as well as rs13306561 T>C variants in MTHFR significantly decreased the risk of ALL in our population. While, no significant association between the MTHFR C677T (rs1801133) variant and ALL was found.21-24  Whereas, few groups have reported that MTHFR C677T variant decreased25 or increased the risk of ALL.12 Our results showed the same findings regarding the role of MTHFR A1298C in the decrease of the ALL risk.12 In contrast to our findings, no significant association between MTHFR A1298C polymorphism and ALL has been found.26-29 Several investigations proposed that MTHFR A1298C polymorphism increased the risk of ALL.22,30-31 Recently, a meta-analysis research has been done in a Chinese population and its results suggested that both MTHFR C677T and A1298C polymorphisms may be potential biomarkers for ALL risk.13In agreement with our finding a number of previous studies have found no significant association between MTHFR C677T variant and risk of childhood ALL. Overall, a significant decreased association was found between the MTHFR C677T polymorphism and ALL risk. In contrast, C677T and A1298C variants were significantly associated with an increased risk of ALL in hospital-based studies, which included individuals with unknown ethnicity.14 have done a meta-analysis and showed that MTHFR rs1801131 CC genotype polymorphism significantly increased the risk of myeloid leukemia in Asian populations. In addition, Jiang et al.25 have also indicated that 677TT genotype decreased the risk of ALL in Caucasians in a meta-analysis study .Dong et al.32 have found an overall association between 677T variant genotypes and reduced childhood ALL risk, though they found no significant association between MTHFR A1298C polymorphism and childhood ALL risk.However, no significant association was found between C677T polymorphism and ALL in Asians and others. In another meta-analysis study, Yan et al.30 It has been reported that individuals carrying MTHFR 677T allele may have a higher relative risk of pediatric ALL mortality.33 In a meta-analysis study, no association has been reported between MTHFR A1298C and C677T polymorphisms and pediatric ALL.29 Recently, it has also been shown that the risk of ALL development in children with MTHFR 677 TT genotype decreased in Asian population.34In another meta-analysis, it has been showed that the 677T allele was protective, while A1298C was associated with marginal increase in the risk of childhood ALL.35 investigated the impact of MTHFR C677T and A1298C polymorphisms and the risk of ALL in patients referred to the Iranian Blood Transfusion Organization . In agreement with our finding, they found no significant association between MTHFR C677T variant and the risk of childhood ALL. Interestingly, they also found no association between MTHFR A1298C polymorphism and the risk of ALL. It is noteworthy to mention that the genetic background of our population  was different from their population, which used in their study .Atashrazm et al.Therefore, it can be concluded that the data regarding the impact of MTHFR on childhood ALL are inconsistent. There is no clear reason for the different results in diverse studies. Ethnic, genetic, and/or environmental factors as well as gene-diet interaction may interact in various ways to either decrease or increase the risk of childhood ALL in different areas.Taken together, our findings showed that MTHFR rs1801131 and rs13306561 polymorphisms decreased the risk of ALL in a sample of Iranian population. To further confirm these findings, larger sample sizes with diverse ethnicities are required. The results of the current case-control study showed an association between MTHFR rs1801131 and rs13306561 polymorphisms and risk of ALL in a sample of Iranian population.


IL1B -511 C>T and IL1RN 86bp VNTR in several of the constructed genetic models (p range = 0.01–0.044 and 0.005–0.034 respectively). The presence of >1, ‘T’ (minor) allele of IL1B -511 C>T in a genotype seemed to provide protection against CAD , while the presence of >1, ‘C’ (major) allele seemed to increase the risk of CAD . The minor allele  of IL1RN 86bp VNTR and its homozygous genotype (2/2 genotype) also seemed to carry an increased risk for CAD . On the other hand, several haplotype combinations constructed out of IL1B and IL1RN gene variants clearly showed statistically significant associations with CAD (p<0.0045). Our meta-analysis was conducted for 8 previously assessed IL-1 SNPs. We included 53 different studies which involved a total sample of 26,210 . Our pooled results concurred with the findings of our case-control study and was not able to deduce any statistically significant associations for any of the 8 studied SNPs (p>0.05). Subgroup analysis, however, yielded interesting results, where significant differences in association statuses were seen for IL1A +4845 G>T, IL1B -511 C>T, IL1RN 86bp VNTR and IL1RN +8006 T>C for select ancestral subgroups. The hints of associations deduced for subjects belonging to NIA in our case-control study for both IL1B -511 C>T and IL1RN 86bp VNTR were duly validated vide significant p values seen for NIA in all three genetic models . On the other hand, Mixed Ancestry (MA) subgroup carrying IL1B -511 C>T, IL1RN 86bp VNTR or IL1RN +8006 T>C polymorphisms seemed to enjoy significant protection against CAD. A few other ancestral subgroups also demonstrated significant associations for a few of the studied SNPs vide one of the three genetic models. Clinical interpretation of derived results is however recommended.Several researchers have reported significant association of numerous single nucleotide polymorphisms (SNPs) residing in the interleukin-1 (IL-1) gene cluster with coronary artery disease (CAD). However, their association status amongst North Indian ancestry (NIA) have never been systematically assessed. Despite a published meta-analysis on this subject, their association status worldwide as well as amongst different major ancestral subgroups still remains unclear. We therefore decided to prospectively test the association of 11 IL-1 gene cluster SNPs with CAD, vide a case-control study amongst a cohort of NIA and attempted to validate our results with the help of an updated meta-analysis of all relevant published association studies. Included studies were segregated into ancestral subgroups and association statuses for each subgroup were determined. A total of 323 cases and 400 healthy, age and sex matched controls belonging to NIA were prospectively enrolled and subsequently genotyped for 11 selected IL-1 gene cluster SNPs. Although results for none of the evaluated IL-1 gene cluster SNPs reached the adjusted level of significance (p<0.0045), clear trends of association were seen for  IL1A, IL1B and IL1RN genes. These three genes are encompassed in the interleukin-1 gene cluster which is located within a 430 kb region on chromosome 2 (2q13-21). All these three genes are highly polymorphic entailing several single nucleotide polymorphisms (SNPs); some of them among which have even been implicated with CAD. Over the years, the associations reported have been area specific and sporadic. A meta-analysis of association studies published on the subject in 2012, reported lack of association of 3 SNPs  residing in IL-1 gene cluster with CAD.. Version 5.3. Copenhagen: The Nordic Cochrane Centre, The Cochrane Collaboration, 2012. The extracted data from all publications were tested using dominant, recessive and allelic genetic models.All calculations in the present meta-analysis were carried out using Review Manager .Bivariate and random or fixed effect models were used for calculating odds ratios (OR’s). For each genetic variant, individual and summary odds ratios and their 95% confidence intervals (CIs) were calculated for all genetic models, using either random (DerSimonian-Laird method) or fixedI2 statistics (I2) and Cochran’s Q statistics (PQ) for each analysis and the study group with a PQ cut-off value of <0.1 was identified as a heterogeneous group. Low, moderate and high heterogeneity was defined according to previously published estimates[I2 values of 25%, 50% and 75% respectively. Criterion for selecting fixed effects for analysis in a group/subgroup was predefined with a cut-off PQ value of 0.10. An attained PQ value of >0.10 in a group/subgroup qualified it for the use of fixed affects for analysis, whereas random effects were used for analysis of groups/subgroups with PQ value of ≤0.10. Subgroup differences were assessed assuming the same cut-off PQ value. Differences yielding a PQ value of ≤0.10 were assumed to be significant, while others (PQ> 0.10) were assumed to be non-significant.Existence of heterogeneity was tested using a Q test. Tests for heterogeneity were performed using Higgins estimates using cuSince most of statistical methods available for detection of publication bias are sensitive to heterogeneity, we used two of the most accepted methods in our updated meta-analysis. Presence of publication bias among the studied groups/subgroups (with ≥3 included studies) was visually detected by the use of Begg’s funnel plots and its Sensitivity analysis was performed in each study group in every genetic model, where we excluded studies one after another and conducted the analysis after each omission. We tested if the results in any of the groups and studied genetic models altered substantially to change the results from non-association to significant association or the other way around (in groups of ≥ 5 studies). Absence of such phenomenon generally indicates the robustness of that meta-analysis.The baseline characteristics of the case and control groups are showcased in IL1B -511 C>T and IL1RN 86bp VNTR certainly showed trends of association with the disease. . Comparisons in co-dominant, dominant and recessive genetic models also did not reach the adjusted statistical significance, but hinted towards the protective nature of the resultant T allele and the TT genotype. The C (major) allele and CC genotype of IL1B -511 C>T thus automatically hinted to be involved with an increased CAD risk. On the other hand, minor allele 2 (240 bp) and genotype 2/2 resulting from IL1RN 86bp VNTR polymorphism showed a trend to be associated with increased risk for CAD . This effect was further inspected, when we performed comparisons after renaming all other alleles  as allele X. Their genotypes were thus renamed as X/X, X/2 and 2/2. Results from comparisons (using the previously described nomenclature) among dominant, recessive and allelic models, although did not reach adjusted statistical significance (p> 0.0045), but showed similar trends of association of genotype 2/2 and allele 2 with CAD.  came out to be as 0.0045. Although, none of the SNPs belonging to IL-1 family, technically provided statistically significant evidence of association (p>0.0045), a couple of them i.e. disease.  The T . Several such haplotypes either imparted additional risk or proved to be a protective factor against CAD. None of the constructed haplotype combinations of IL1A gene variants however showed any hint of association with the disease. . This left us full texts of 28 articles to be assessed for eligibility. Among them 7 papers were excluded for a variety of reasons e.g.: non-availability of complete information (n = 4), genetic distribution amongst controls not conforming to Hardy-Weinberg approximations (n = 2) and published in any other language than English (n = 1). Finally 21 papers (plus 1 present article), with 53 different studies (including 8 from the present manuscript) were included for quantitative synthesis. Since we were not able to find a single relevant and qualifying, published study for 3 nalyzed.  Overall nalyzed.  All inclIL1A gene viz. -889 C>T and +4845 G>T were selected for meta-analysis. We included a total of three studies, including ours’,[I2< 50% and PQ> 0.10). , while the differences did not reach statistical significance in recessive and allelic models (p>0.05). . , S1 Fig p>0.05). , Fig 2IL1B gene, viz. -511 C>T, -5887 C>T, -3954 C>T and -1903 C>T, were selected for the present meta-analysis. Fifteen different studies (including ours’) were included for -511 C>T, with a total sample of 7,429  and the quantitative synthesis was carried out using appropriate effects for analysis.[I2 = 54.5%, PQ = 0.07 and I2 = 51.3%; PQ = 0.10 respectively). . , Fig 3IL1B SNPs (I2< 50% and PQ> 0.10). . –S4 FigsIL1RN gene. Including the present study, a total of 15 different studies[I2 = 79%, PQ = 0.003 in dominant I2 = 62.4%, PQ = 0.05 in recessive and I2 = 83.4%; PQ = 0.0004 in allelic genetic model). . , Fig 4I2 = 85.8%, PQ = 0.0009 and I2 = 76.8%, PQ = 0.01 respectively). . , Fig 5IL1B -511 C>T and IL1B -3954 C>T, polymorphisms, whereas IL1RN 86bp VNTR and IL1RN +8006 T>C polymorphisms. Each point in these plots represents the OR of a study plotted against the standard error (SE) of this aforementioned OR. Studies belonging to different ancestral subgroups have different indicators in these plots and they seem to be generally contained within the inverted cone, indicating no significant publication bias, at least in the pooled analysis. However, Egger’s test which although detected no significant publication bias for most the groups/subgroups, did manage to produce some significant p values in a few groups/subgroups. Subgrouping on the basis of ancestry did successfully tone down the observed significant p values seen in the pooled analysis, indicating heterogeneity as the culprit. Although, all the groups with significant Egger’s p value were small and the culprit could have been inherent heterogeneity, no significant Egger’s p values were detected in the groups/subgroups which were significantly associated with the disease.  which had ≥3 included studies using two statistical methods viz. Begg’s funnel plot test and Eggedisease.  FigsSensitivity analysis was performed in each study group, where we excluded studies one after another and conducted the analysis after each omission (in groups of ≥ 5 studies). Only three SNPs viz. B-511 C>T, B-3954 C>T, RN86bp VNTR qualified for the same. The summary effects deduced for these SNP’s  did not alter significantly i.e. from non-association to significant association or the other way around, indicating the robustness of the present meta-analysis.et al.[The present association study is the largest so far ever conducted amongst ethnic North Indians both in terms of the sample size assessed and the number of IL-1 gene cluster variants studied. Power analysis was also conducted using the generated data for all 11 SNPs. Our power analysis indicated that the sample size of our case-control study was sufficient to support the derived results . Our updated meta-analysis is the most comprehensive on the subject presented so far. It has yielded novel yet reproducible insight into the association statuses of IL-1 gene cluster variants with CAD amongst diverse ancestral subgroups. The first and foremost merit of the present meta-analysis is that it has been conducted ensuring strict adherence to all contemporary recommendations and guidelines. It can be considered far superior to the only published meta-analysis on this subject, both in et al.; a mistaIL1A gene SNPs (i.e. -889 C>T and +4845 G>T) with CAD. Neither of the possible haplotypes constructed out of the two SNPs showed any significant association. Neither pooled nor our subgroup analysis deduced any significant associations for -889 C>T. Such result was expected, as none of the evidence published so far indicated any hint of association. Non association with restenosis after percutaneous transluminal coronary angioplasty (PTCA) has also been reported for -889 C>T which probably testifies its lack of association with the disease.[Our case-control study suggested lack of association of both the studied et al.[Only one study so far, (amongst Middle Easterners) has reported significant association of +4845 G>T with CAD in a relatively small sample size of 200 (100 cases and 100 controls). Its contet al. lacks biIL1A SNPs. This explains their lack of association seen in our case-control study and the present meta-analysis.Absence of any relevant and statistically significant published data indicates non-functional nature of both these IL1B SNPs , in the present case-control study yielded desired statistical significance to confirm association, distinct trends suggesting possible association were seen for -511 C>T. Congruent with the results of our case-control study, subgroup analysis for -511 C>T hinted associations for certain ancestral subgroups (AA), and even attested the same for the NIA subgroup, vide all three genetic models. Results of our case-control study as well as our meta-analysis suggested that C to T substitution at position 511bp of the IL1B gene might act as protective factor against CAD, at least amongst subjects of North Indian Ancestry. Over the years most association studies amongst subjects belonging varied ancestries,[IL1B -511 genotypes with CRP values has already been established,[Although none of the association tests for all 4 cestries,, 32–37 acestries, have faicestries,, 33 Thisablished, the protablished,, 49 Furtablished, They fouablished, The releablished, Mononuclablished, This speablished, confirmsIL1B SNPs have been extensively investigated with respect to CAD/MI, yielding negative results.[Lack of associations for -1903 C>T, -3954 C>T and -5887 C>T with CAD, as seen in our case-control study were attested by both the pooled as well as subgroup results of our meta-analysis. Over the last decade, these aforementioned  results.–36, 38 L results.. Since wIL1B haplotypes that might impart diverse effects on the outcome. The underlying biochemical mechanisms behind these diverse effects are unknown and warrant functional studies.Our case-control study also identified several IL1RN gene polymorphisms  in our present case-control study. Association status tested for IL1RN 86bp VNTR with CAD, both in our case-control study as well as in the pooled results our meta-analysis also did not reach the desired statistical significance. These results are in concordance with results of only published meta-analysis on this subject.[et al.,[No hint of association was seen when we compared genotypic and allelic frequencies in cases with that in controls for 4 out of 5 selected  subject., 43 it a subject. Latkovsk.[et al., reported.[et al.,, 43 Gene.[et al., it is poOur meta-analysis although validated the lack of association for +8006 T>C in the pooled analysis, but discovered significant yet contrasting association trends in a couple of ancestral subgroups (p<0.05). While the dominant model of the EA subgroup suggested this SNP as a risk factor for CAD, the dominant and allelic model for the MA subgroup suggested it to be protective factor. Published literature does not explain the underlying biological basis of such contrasting associations and thus probably requires further investigation.IL1A +4845 G>T, IL1B -511 C>T, IL1RN +8006 T>C and IL1RN 86bp VNTR) which duly qualifies to be a highlight of the present meta-analysis.Our pooled result also concurs with the CARDIoGRAMplusC4D GWAS data, which incidentally is the most comprehensive meta-analysis of published GWAS studies on the subject. Severe aIL1B-511C>T and ILIRN 86bp VNTR) have previously been thoroughly (if not conclusively) investigated with respect to CAD. Also, their biological plausibility has at least been previously indicated, if not established. The fact that this aforementioned plausibility is not yet conclusively established may qualify to be a primary limitation of our study. Also since very few published reports are available from our region, and since it’s a land of significant genetic diversity, we are not sure if our results would hold true in other population pockets of India. More association studies among cohorts belonging to North Indian ancestry are thus warranted in order to establish/negate these seen associations. Ideally, association among serum levels of interleukins with their genotypes should also have been investigated in the present case-control study, however arranging the logistics of same in a multicentric study, like ours’ was difficult, and this qualifies to be another major limitation. Possibility of errors in genotyping, presence of selection bias and risk of inadequate sample size amongst different studies could easily qualify to be listed as limitations in every meta-analysis like ours’. Any meta-analyses of association studies cannot inspect interference of linkage disequilibrium and thus cannot measure its overall effect on the seen associations. Although, no significant publication bias was detected amongst our study groups, the role of existing publication bias however cannot be completely ruled out which also qualifies to be listed as another small limitation in our present meta-analysis.Every genetic association study and their meta-analyses for a particular disease/condition have their potential limitations. The clinical utility of association studies like ours’ is always in question and the derived significant p values should be always be accompanied by biological plausibility of the seen association. In the present case-control study the two SNPs hinting association with the disease  allele seemed to increase the risk of CAD in our study cohort. Conversely allele 2  and genotype X/2 of IL1RN 86bp VNTR polymorphism seemed to increase an individual’s risk for CAD. Several haplotype combinations constructed out of studied SNPs belonging to IL1B and IL1RN genes also showed varied statistically significant associations with CAD.Although our association study amongst NIA failed to conclusively attest to statistically significant associations for all 11 studied IL-1 gene cluster SNPs with CAD, we found hints suggesting possible association for at least a couple of them (viz. B -511 T>C and RN 86bp VNTR). While the presence of >1, T (minor) allele of IL1A +4845 G>T, IL1B -511 C>T, IL1RN 86bp VNTR and IL1RN +8006 T>C. The hints of associations deduced for subjects belonging to NIA in our case-control study for both IL1B -511 C>T and IL1RN 86bp VNTR were duly validated vide significant p values seen in our subgroup analysis. On the other hand, MA subgroup carrying IL1B -511 C>T, IL1RN 86bp VNTR and IL1RN +8006 T>C polymorphisms seemed to enjoy significant protection against CAD. A few other ancestral groups also showed significant associations vide any one of the three constructed genetic models for a few of the studied SNPs, however these associations did not seem conclusive and do require further validation. Although the practical utility of such results are always questionable, the generated information however has the potential to be used after clinical interpretation.Congruent with the results of our association study, the pooled results of our meta-analysis also did not show any statistically significant associations between studied IL-1 gene cluster polymorphisms and CAD. However, significant differences in association statuses amongst studied ancestral groups were seen for S1 Checklist(DOC)Click here for additional data file.S2 Checklist(DOC)Click here for additional data file.S1 Data(ZIP)Click here for additional data file.S1 FigTT+CT vs. CC); Panel B: Effect size estimation using recessive genetic model (TT vs. CT+CC); Panel C: Effect size estimation using allelic genetic model (Allele T vs. Allele C). Effect size estimates for dominant, recessive and allelic genetic models  were obtained using fixed effects for analysis.Panel A: Effect size estimation using dominant genetic model ((TIF)Click here for additional data file.S2 FigTT+CT vs. CC); Panel B: Effect size estimation using recessive genetic model (TT vs. CT+CC); Panel C: Effect size estimation using allelic genetic model (Allele T vs. Allele C). Effect size estimates for dominant, recessive and allelic genetic models  were obtained using fixed effects for analysis.Panel A: Effect size estimation using dominant genetic model ((TIF)Click here for additional data file.S3 FigTT+CT vs. CC); Panel B: Effect size estimation using recessive genetic model (TT vs. CT+CC); Panel C: Effect size estimation using allelic genetic model (Allele T vs. Allele C). Effect size estimates for dominant, recessive and allelic genetic models  were obtained using fixed effects for analysis.Panel A: Effect size estimation using dominant genetic model ((TIF)Click here for additional data file.S4 FigTT+CT vs. CC); Panel B: Effect size estimation using recessive genetic model (TT vs. CT+CC); Panel C: Effect size estimation using allelic genetic model (Allele T vs. Allele C). Effect size estimates for dominant, recessive and allelic genetic models  were obtained using fixed effects for analysis.Panel A: Effect size estimation using dominant genetic model ((TIF)Click here for additional data file.S5 FigIL1B -511 C>T polymorphism. Panel B: Begg’s funnel plot with Egger’s estimates for recessive genetic model of IL1B -511 C>T polymorphism. Panel C: Begg’s funnel plot with Egger’s estimates for allelic genetic model of IL1B -511 C>T polymorphism. Panel D: Begg’s funnel plot with Egger’s estimates for dominant genetic model of IL1B -3954 C>T polymorphism. Panel E: Begg’s funnel plot with Egger’s estimates for recessive genetic model of IL1B -3954 C>T polymorphism. Panel F: Begg’s funnel plot with Egger’s estimates for allelic genetic model of IL1B -3954 C>T polymorphism. Abbreviations- EA: European Ancestry; MEA: Middle Eastern Ancestry.Each point in each figure represents OR of a study plotted against the standard error (SE) its OR. Different indicators of the studies belonging to each ancestral group are used in these plots. Panel A: Begg’s funnel plot with Egger’s estimates for dominant genetic model of (TIF)Click here for additional data file.S6 FigIL1RN 86bp VNTR polymorphism. Panel B: Begg’s funnel plot with Egger’s estimates for recessive genetic model of IL1RN 86bp VNTR polymorphism. Panel C: Begg’s funnel plot with Egger’s estimates for allelic genetic model of IL1RN 86bp VNTR polymorphism. Panel D: Begg’s funnel plot with Egger’s estimates for dominant genetic model of IL1RN +8006 T>C polymorphism. Panel E: Begg’s funnel plot with Egger’s estimates for recessive genetic model of IL1RN +8006 T>C polymorphism. Panel F: Begg’s funnel plot with Egger’s estimates for allelic genetic model of IL1RN +8006 T>C polymorphism. Abbreviations- EA: European Ancestry; MEA: Middle Eastern Ancestry.Each point in each figure represents OR of a study plotted against the standard error (SE) its OR. Different indicators of the studies belonging to each ancestral group are used in these plots. Panel A: Begg’s funnel plot with Egger’s estimates for dominant genetic model of (TIF)Click here for additional data file.S1 File(DOC)Click here for additional data file.S1 Table(DOC)Click here for additional data file.S2 Table(DOC)Click here for additional data file.S3 Table(DOC)Click here for additional data file.S4 Table(DOC)Click here for additional data file.


Several single nucleotide polymorphisms (SNPs) in an α-neuronal nicotinic acetylcholine receptor subunit (CHRNA3/5) were identified to be associated with chronic obstructive pulmonary disease (COPD) in a study based on a Norwegian population. However, results from subsequent studies have been controversial, particularly in studies recruiting Asians. In the present study, we conducted a comprehensive search and meta-analyses to identify susceptibility SNPs for COPD in the CHRNA3/5 locus.A comprehensive literature search was conducted to find studies that have reported an association between SNPs in the CHRNA3/5 locus and COPD risk. Pooled odds ratios (ORs) with 95% confidence intervals (CIs) for each SNP were calculated with the major allele or genotype as the reference group. The influence of individual studies on pooled measures was assessed, in addition to publication bias.A total of 12 articles with 14 eligible studies were included in this analysis. Association between 4 SNPs in the CHRNA3/5 locus and COPD was evaluated and included rs1051730, rs8034191, rs6495309, and rs16969968. Significant associations between the 4 SNPs and COPD were identified under allele  and genotype models. Subgroup analysis conducted for rs1051730 showed a significant association between this SNP and COPD risk in non-Asians , but not Asians . Rs1051730 and rs6495309 were also significantly associated with COPD after adjusting for multiple variables, including age and smoking status.Our results indicate that 4 SNPs in the CHRNA3/5 locus are associated with COPD risk. Rs1051730 was particularly associated with COPD in non-Asians, but its role in Asians still needs to be verified. Additional studies will be necessary to assess the effect of rs6495309 on COPD. Although rs1051730 and rs6495309 were shown to be independent risk factors for COPD, validation studies should be performed. Tobacco smoking is a major risk factor for development of chronic obstructive pulmonary disease (COPD) A comprehensive search was performed in several databases, including Pubmed, ISI Web of Science, CNKI , CBM , and Wangfang Data for potentially relevant studies published in English or Chinese. An upper date limit of Dec 1, 2013 was applied, and a lower date limit was not specified. A combination of the following Medical Subject Headings (MeSH) terms and key words were used: “pulmonary disease, chronic obstructive”, “lung disease, obstructive”, “COPD”, “chronic obstructive pulmonary disease”, “receptors, nicotinic”, “nicotinic receptor subunit”, “nAChR”, “CHRNA”, and “15q25”. Bibliographies of relevant articles were manually reviewed to identify additional studies that were not captured in the key word search.Studies that assessed the association between any SNP in the CHRNA3/5 locus and COPD risk and reported genotype count, allele frequency, or odds ratios (ORs) with 95% confidence intervals (CIs) were included in this study. Case-control, cohort, or population-based studies were eligible, although studies without a control group or based on family or sibling pairs were excluded. The study with the largest sample size was included when 2 or more studies contained the same data set. All potentially available articles were independently identified by 2 reviewers to determine study eligibility. Consensus with a third reviewer was used to resolve disagreements regarding eligibility.Extracted data included the following: name of first author, publication year, country in which the study was conducted, ethnicity of the study population, type of study design, source of control subjects, mean age, male percentage, smoking status, genotyping methods, definition of COPD, matched methods, total number of cases and controls, genotype count, allele frequency, and ORs with 95%CIs adjusted for multiple variables.The Newcastle-Ottawa quality assessment scale was used to evaluate the quality of each study. Assessment for case-control studies was based on 3 major components, including selection of cases and controls, comparability of cases and controls, and exposure. Prospective cohort studies were evaluated on another scale consisting of 3 components, including selection of exposed and non-exposed populations, comparability of cohorts, and outcome. The assessment was conducted independently by 2 reviewers, and disagreements were resolved by consensus with a third reviewer.2 statistic was used to evaluate heterogeneity among studies, which describes the proportion of total variation attributable to between-study heterogeneity. I2>50% was considered significant statistical heterogeneity, and pooled ORs were calculated using the DerSimonian and Laird random effect model. For I2<50%, the fixed effect model was used. If needed, subgroup analyses were conducted to explore the source of heterogeneity. One-way sensitivity analyses in which a single study was omitted were employed to assess robustness of the pooled estimators in response to removal of an individual study. When the point estimate of the omitted analysis was outside the 95%CI of the pooled measures, the individual study was suspected of excessive influence. Begg's test was conducted to estimate publication bias and displayed as a funnel plot. All statistical analyses were performed with STATA version 9.2 . All reported probabilities were two-tailed, with P<0.05 considered statistically significant.SNPs reported by more than 3 independent studies underwent further meta-analysis. Fisher's exact test was used to analyze deviations from Hardy-Weinberg equilibrium (HWE) for each genotype distribution in the control group. The inverse variance-weighted mean was used to calculate pooled ORs with 95%CIs for each SNP to represent association with risk of COPD under allele, homozygote, and heterozygote genotype models. The major allele or genotype was used as the reference group. The IA study identification flow diagram is shown in 2 was 14.0%, and a fixed effect model was used. Allele distribution was significantly different between the groups, with an OR of 1.14 for the T vs C allele  (As shown in = 0.000) . Conside2 was 0.0%, and the fixed effect model was used to calculate association of the C allele with COPD risk compared with the T allele. The C allele was significantly associated with the risk of COPD   .2 = 50.4%), and a random effect model was used to calculate association of the C allele with COPD risk compared with the T allele. The C allele was associated with a significantly increased risk of COPD   .2 was 0.0%, and the OR for the A allele regarding risk of COPD compared with the G allele was 1.27   .ORs with 95%CIs for heterozygote and homozygote genotypes were also calculated . Under bTo adjust for other confounding factors, such as smoking, age, and gender, we conducted meta-analyses for eligible studies that reported ORs with 95%CIs adjusted for multiple variables. For rs1051730 and rs6495309, 3 studies were included in the meta-analyses. Pooled ORs with 95%CIs were 1.11 (1.01–1.20) and 1.31 (1.08–1.54), respectively .No strong influence of any single study was detected after it was omitted, except for the rs1051730 heterozygote genotype . After tIn 2011, Zhang et al conducted a meta-analysis to assess the effect of rs1051730 in the CHRNA3/5 locus on COPD While searching for eligible studies, we found 17 SNPs in the CHRNA3/5 locus that were reported to be associated with COPD, including rs1051730, rs8034191, rs6495309, rs16969968, rs578776, rs12910984, rs660652, rs1317286, rs569207, rs12914008, rs931794, rs938682, rs17486278, rs7180002, rs951266, rs684513, and rs588765. However, only the former 4 SNPs were studied in at least 5 individual studies. For each SNP included in this study, we compared allele and genotype distribution between cases and controls. Our results showed that the T allele in rs1051730 was associated with COPD risk, which was similar for both homozygote and heterozygote genotypes. This conclusion was consistent with the studies by Pillai et al and Zhang et al. In the present study, we found the polymorphism frequency was far lower in Asians. The minor allele frequency (MAF) in the non-Asian subgroup was greater than 0.48. The largest MAF was 0.031 from Guo's study in an Asian subgroup. We further evaluated the association between rs1051730 and COPD in different ethnic subgroups and showed that the T allele increased COPD risk, particularly in non-Asians. However, based on the results, rs1051730 does not appear to be associated with COPD in Asians. Because only 3 of the studies included an Asian subgroup and the MAF was lower than in non-Asians, additional studies are still needed to confirm the association between this polymorphism and COPD risk in Asians.Rs8034191 is another susceptibility SNP identified in the study of Pillai et al. Our results showed that the rs8034191 C allele was associated with COPD risk under any of the inheritance models, including allele, homozygote, and heterozygote genotypes. Given that there was only one study based on an Asian population, subgroup analysis for different ethnicities was not conducted. Similar to rs1051730, the MAF of this SNP was 0.031 from Zhou's study based on Chinese subjects, while the lowest MAF in the studies with non-Asian participants was 0.51 , which was based on Caucasians. We suspect that there is a stronger association between rs8034191 and COPD risk in non-Asians than Asians, which should be confirmed through additional studies.Based on the results of rs6495309 and rs16969968, an association between SNPs and COPD risk was detected. The rs6495309 C allele and rs16969968 A allele were considered risk alleles that increased the risk of COPD. The studies included in meta-analyses for rs6495309 were mainly based on Chinese or Korean populations, and none of the studies reported an association between rs6495309 and COPD in Caucasians or non-Asians, including the GWA studies, indicating that this SNP is likely an important susceptibility variant for COPD in Asians.We tried to detect whether these SNPs in the CHRNA3/5 locus were associated with COPD risk by direct or indirect influence. We pooled the ORs with 95%CIs adjusted for multiple variables, including smoking status and pack-years of smoking. Eligible data were found in the rs1051730 heterozygote genotype and rs6495309 homozygote genotype. Although the number of included studies was small, the results still suggest that these variants are associated with COPD risk, independent of smoking behavior or nicotine dependence. This conclusion needs to be further confirmed with additional studies based on never smokers.In sensitivity analyses, the study by Kaur-Knudsen et al had a strong influence on pooled measures under an inheritance model of rs1051730, with the pooled OR increasing after omitting this individual study. However, the pooled OR was still statistically significant at P = 0.05 and the results after omitting the study still confirmed an association between the polymorphism and COPD. Publication bias was detected under some inheritance models of rs6495309. In this meta-analysis, only 4 studies on the association of rs6495309 with COPD were included, and the bias could be caused by the small number of studies. According to our results, rs6495309 is associated with COPD, but future studies will need to confirm this conclusion regarding publication bias.Based on our study, 4 SNPs in the CHRNA3/5 locus were associated with COPD risk. Among the 4 SNPs, the rs1051730 T allele was considered a risk allele, particularly in non-Asians. However, the effect of both rs1051730 in Asians and rs6495309 on COPD risk needs to be confirmed by further studies. Although rs1051730 and rs6495309 were found to be independent risk factors for COPD, this effect will need to be confirmed.Table S1General characteristics and quality assessment of each study.(DOC)Click here for additional data file.Table S2Distribution of the SNP alleles and genotypes.(DOC)Click here for additional data file.Checklist S1PRISMA meta-analysis checklist.(DOCX)Click here for additional data file.


Q = 82.48, P < 0.001, I2 = 79.4%). We conducted stratified and meta-regression analyses to identify the origin of the heterogeneity among the studies, and a Galbraith plot was generated to graphically assess the sources of heterogeneity. This meta-analysis provides a robust estimate of the positive association between maternal folate supplementation and a decreased risk of CHDs.Epidemiological studies have reported conflicting results regarding the association between maternal folic acid supplementation and the risk of congenital heart defects (CHDs). However, a meta-analysis of the association between maternal folic acid supplementation and CHDs in offspring has not been conducted. We searched the MEDLINE and EMBASE databases for articles cataloged between their inceptions and October 10, 2014 and identified relevant published studies that assessed the association between maternal folate supplementation and the risk of CHDs. Study-specific relative risk estimates were pooled using random-effects or fixed-effects models. Out of the 1,606 articles found in our initial literature searches, a total of 1 randomized controlled trial, 1 cohort study, and 16 case-control studies were included in our final meta-analysis. The overall results of this meta-analysis provide evidence that maternal folate supplementation is associated with a significantly decreased risk of CHDs . Statistically significant heterogeneity was detected ( Congenital heart defects (CHDs) are the most common congenital malformations, affecting nearly 1% of live births worldwideSeveral chromosomal anomalies, certain maternal illnesses, and prenatal exposures to specific therapeutic drugs are recognized risk factors. It is difficult to establish the role of a single factor because the cause of a defect is believed to be multifactorial in many cases; for example, some cases may result from a combination of environmental teratogens with genetic and chromosomal abnormalitiesMore than a decade ago, the preventive effects of maternal folate supplementation on the occurrence and recurrence of neural tube defects was documented in several studies5An increasing number of studies to date have focused on the association between maternal folic acid supplementation and the incidence of CHDs; however, the results have been ambiguous, perhaps due to inadequate sample sizes. Thus, we conducted a meta-analysis to quantitatively assess the effect of maternal folic acid supplementation on the risk of CHDs.Our literature search strategy generated 1,606 citations. Of these, 18 were used in the final analysis, representing 18,500 incident cases . All of 101112131415161718192021222324Q = 82.48, P < 0.001, I2 = 79.4%), with no publication bias , after omission of the study by Malik et al.Q = 72.29, P = 0.000, I2 = 77.9%), after omission of the study by Li et al.The overall results of this meta-analysis provided evidence for a significant decrease in the risk of CHDs with maternal folate supplementation . We further created a Galbraith plot to graphically assess the sources of heterogeneity (I2 = 31.9%); however, the corresponding pooled RRs were not materially altered .To clarify the sources of heterogeneity, we conducted a sensitivity analysis. However, ogeneity . A totalTo our knowledge, this is the first quantitative meta-analysis to evaluate the association between maternal folate supplementation and the risk of CHDs. Overall, the findings of our meta-analysis suggest that maternal folate supplementation is significantly associated with a decreased risk of CHDs . Moreover, these results were consistent across most of the subgroup analyses .Although the specific biological mechanisms underlying the relationship between maternal folate supplementation and the risk of CHDs remain to be determined, some relevant evidence has been published to date. It has been hypothesized that impaired folate and/or homocysteine metabolism interferes with the development of the heart, possibly by affecting neural crest cells. Methylenetetrahydrofolate reductase (MTHFR), which is a critical folate-metabolizing enzyme, plays an important role in processing amino acids. A C→T substitution is commonly found at position 677 in the MTHFR enzyme and results in a substitution of valine for alanine; this substitution causes impaired folate binding and reduced activity of the MTHFR enzyme3033Although the potential role of folic acid in the prevention of neural tube defects was reported as early as 1980, public health campaigns have resulted in preconception supplementation in only one-third of pregnant womenSome limitations of our study must be taken into account. First, 1 randomized controlled trial, 1 cohort study, and 16 case-control studies were included in our meta-analysis, and we extracted our raw data primarily from several case-control studies, which were susceptible to selection and information biases. In addition, our meta-analysis was limited to studies published in English, so our results may have been affected by a lack of data from studies published in other languages. Thus, our conclusions must be considered carefully. Second, although no evidence of publication bias was found, heterogeneity was among the studies included in the analysis, and this heterogeneity may affect the interpretation of the overall results. In this study, we conducted sensitivity analyses to explore the sources of this heterogeneity by removing one study at a time from our pooled analysis. However, heterogeneity could not be fully removed by the exclusion of any individual study. Moreover, study results may vary with geographical regions, publication periods, sample sizes, CHD subtypes, and other risk factors. Thus, we performed meta-regression and subgroup analyses to further investigate the sources of the heterogeneity that we observed. We found that the heterogeneity in the study results could partly be attributed to the geographical region in which the studies that we examined were conducted. We also created a Galbraith plot to assess the heterogeneity that we observed and to identify potential outlying studies. A total of 8 studies were identified to be the primary contributors to the heterogeneity in the analysis. After excluding these outlying studies, the heterogeneity was effectively removed, whereas the corresponding pooled RRs were not materially altered, indicating that the overall results regarding maternal folate supplementation were statistically stable.Our study has several important strengths. First, to our knowledge, this is the first meta-analysis to report an association between maternal folate supplementation and the risk of CHDs. Moreover, our literature search was conducted using multiple databases, and the references from the retrieved articles were carefully examined to find any additional studies that may have been of interest. Thus, our study included data for 18,500 cases, which is enough to have sufficient statistical power to investigate the potential association between maternal folate supplementation and the risk of CHDs. Another strength of our study is that although heterogeneity exists in our meta-analysis, we conducted a number of sensitivity, subgroup, and Galbraith plot analyses and found that our results were stable.In summary, this study provides evidence that maternal folate supplementation is positively associated with a decreased risk of CHDs. However, more prospective studies, particularly in developing countries, are needed to further investigate the association between maternal folate supplementation and the risk of CHDs, especially with regard to the different subtypes of CHDs.A computerized literature search was conducted by two independent investigators (Feng and Tong) using the MEDLINE and EMBASE databases to find articles catalogs from the inceptions of these databases through October 10, 2014. We searched for relevant studies using the following search strategy: (“Multivitamins” OR “Vitamin” OR “Folate” OR “Folic Acid”) AND  AND . In addition, we searched for studies that investigated a broad range of environmental teratogens and CHDs and examined the relevant references and review articles that were found. We were thus able to identify relevant information found in other related studies. We followed published quality standards for conducting and reporting meta-analysesWe selected articles that (1) were original epidemiologic studies , (2) examined the association between periconceptional folic acid use and either CHDs overall or any one of the CHD subtypes in infants, (3) were published in the English language, (4) reported RRs  and associated 95% confidence intervals (CIs) or provided raw data from which these measures could be calculated, and (5) defined CHDs or one of the CHD subtypes as an outcome. Articles that reported results from more than one population were considered to consist of separate studies, with 1 study for each population investigated. When multiple articles were found to examine the same study, we included in our study the article with the most applicable information and the largest number of cases. We excluded non-peer-reviewed articles, experimental animal studies, ecological assessments, correlation studies, and mechanistic studies.Data extraction was conducted separately by two reviewers (Feng and Wang) working independently. If differences of opinion arose, these were resolved by discussion between the two reviewers. The studies that met the inclusion criteria were reviewed to retrieve relevant information. Relevant information included author names, the year of publication, the geographic region in which the study was conducted, the period in which data were collected, the study design, the sample size, case classification information, exposure and outcome assessments, adjusted estimates and their corresponding 95% CIs, and confounding factors that were controlled for by matching cases or adjustments in the data analysis. When no adjusted estimates were available, we extracted a crude estimate. If no estimate of relative risk was provided in a given study, we recalculated odds ratios or risk ratios and 95% CIs from the raw data presented in the study using standard equations.To assess study quality, we used a 9-star system based on the Newcastle-Ottawa ScaleQ and I2 statistics were used to test for heterogeneity among studiesP < 0.05 or I2≧56%), a random-effects model was used, which provided a more appropriate summary estimate for heterogeneous study-specific estimates. If the study revealed no evidence of heterogeneity, a fixed-effects analysis was conducted, and an inverse variance weighting was applied to calculate summary RR estimatesWe used study-specific relative risks as summary statistics for the association between maternal folate supplementation and CHD risk. To simplify the procedure, an RR was used to represent all reported study-specific results from cohort studies, and an OR was used to represent results from case-control studies. Cochran's P-value less than 0.05 for the meta-regression analysis was considered to indicate a significant difference between subgroups. Finally, we conducted sensitivity analyses to explore whether a specific study strongly influenced the results, by excluding one study at a time.We conducted subgroup analyses based on study design , geographical region , number of cases , publication period , primary focus of the study , and study quality . We evaluated heterogeneity between subgroups by meta-regression analysis. A P-value of <0.05. All statistical analyses were performed using STATA software .Publication bias was assessed via visual inspection of a funnel plot for asymmetry using both Egger's linear regressionX.M.M. conceived and designed the experiments. Y.F., S.W. and X.M.M. performed the experiments. Y.F., S.W., R.S.C., X.T. and Z.Y.W. analyzed the data. R.S.C. and Z.Y.W. contributed software, hardware and analysis tools. Y.F. and S.W. wrote the paper. X.M.M. will provide access to the full-text article.Supplementary Figure S1


Objective: In the present study, we evaluated the association of rs662799 variant of the APOA5 gene with Metabolic syndrome (MetS) in a sample of children and adolescents from Isfahan.Methods: This case control study comprised 50 cases of MetS and 50 controls. Mismatched polymerase chain reaction–restriction fragment length polymorphism (mPCR-RFLP) was used to genotype -1131T>C polymorphism.Findings: No significant association was documented for APOA5 genotypes with the measured laboratory parameters for CC, CT, and TT genotypes in the two groups studied. By logistic regression using a dominant model, the odds ratio (95% confidence interval0 for the MetS was 0.38 (0.139–1.0350 and 0.29 (0.08–1.071 for the unadjusted and adjusted models, respectively.Conclusion: This study suggests that among studied children and adolescents, -1131T>C polymorphism in the APOA5 gene may not be a major contributor to the MetS risk. The prevalence of the syndrome is 1-2% in Iranian children and adolescents, much higher than that reported for other ethnicities[2-4]. It markedly increases the risk of developing cardiovascular diseases and type 2 diabetes[5]. According to the ATPIII criteria, MetS can be diagnosed based on hypertension, central obesity, insulin resistance and dyslipidemia. However, the most frequent components of the MetS in Iranian children and adolescents are low high-density lipoprotein cholesterol (HDL-C) and high triglycerides (TG)[2].Metabolic syndrome (MetS) is a prevalent complex disorder consisting of concurrent metabolic abnormalitiesAPOA5). The mature APOA5 protein expresses exclusively in the liver and secretes into the plasma to modulate TG metabolism[6]. Among the genetic variants associated with the development of MetS, there is a naturally occurring variant (-1131T>C) in the promoter region of apolipoprotein A5 gene  variant of [2]. In brief, a MetS subject fulfilled at least three of the following components: waist circumference >75th percentile for age and gender in the studied population; fasting TG ≥100mg/dl; HDL-C <50 mg/dl (except in 15-19 year boys in whom the cut off was <45 mg/dl); systolic blood pressure/diastolic blood pressure >90th percentile recommended cut off points by the National Heart, Lung and Blood institute for gender, age and height[14]; fasting blood sugar (FBS) ≥100 mg/dl[2]. This study was approved by the Ethic Committee of Isfahan University. Informed consent was obtained from parents. The experimental design was approved by the Ethics Committee of Shahrekord University.A total of 50 cases of MetS and 50 controls were recruited to this study. Controls had normal weight and were healthy looking, without any clinical, laboratory or history records for MetS, diabetes or cardiovascular disorders. To diagnose the individuals with MetS, we used the modified ATPIII definition[7]. The amplified fragment possessed an obligatory cleavage site for TruI restriction endonuclease to check if the digestion has occurred. Thermal conditions of the PCR cycles were as follow: initial denaturation at 95°C followed by 35 amplification cycles consisting of denaturation at 95°C for 40 sec; annealing at 61°C for 40 sec; extension at 72°C for 40 sec and a final extension at 72°C for 10 min. 10 μl of the PCR product was digested with 5 U of TruI enzyme. According to the digestion patterns, three genotypes were determined: TT genotype resulted in 21, 108 and 267 bp fragments, TC genotype created 21, 108, 267 and 288 bp products and CC genotype produced 108 and 288 bp fragments. Blood specimens were collected in EDTA-treated tubes and were stored at –70°C for further analysis. Extraction of DNA from total blood was performed using the Diatom DNA Prep 100 kit . mPCR-RFLP was used to genotype T>C polymorphism with primers described elsewhereP values less than 0.05 were considered significant. A Chi-square statistic was calculated to compare the frequencies of genotypes and alleles. One way ANOVA was conducted to evaluate any differences between different groups, regarding the levels of biochemical factors and the distribution of different genotypes. Logistic regression analysis was performed to derive the odds ratios. Statistical analyses were done with SPSS software (version 20.0). FindingsP<0.01). The major relevant clinical and biochemical characteristics of the study participants are presented in APOA5 genotypes with any of laboratory parameters. Genotype and allele frequencies are shown in P=0.05. The Odds ratio adjusted for sex, age, HDL-C and TC was also calculated  and no significant differences between adjusted and unadjusted models were found (re found .. Our finding was in agreement with Mattei 's[19] results which showed no association between TG levels and the -1131T>C.We did not find an association between the -1131C polymorphism and elevated triglyceride levels. A number of studies have reported a significant association between triglyceride levels and the -1131T>C variant in MetS adults and obese children[20]. Moreover, no association was found between the -1131T>C and MetS in our study. Significant correlation of -1131T>C APOA5 variant with MetS in adults has been indicated in several[8-13] but not all previous studies. The findings of the current study are consistent with a recent meta-analysis[21] which showed that C allele carriers of -1131T>C had overall a significantly higher risk of MetS. After performing subgroup analysis according to ethnicity, the association was only significant in Asians, but not in white populations.This inconsistency may be due to ethnicity influencesAPOA5 SNPs and haplotypes in Iranian children and adolescents.We showed that -1131T>C variant was neither associated with triglyceride levels nor MetS. Our results call for further studies to explore the effect of other S.G. Fatemi: Doing experiments, analysis of data, drafting the article. M. Emadi-Baygi: Conception, designing the study, analysis and interpretation of data the article. P. Nikpour: Conception, designing the study, analysis and interpretation of data and drafting and revising the article. R. Kelishadi: Sample collection, designing the study, analysis and interpretation of data the article. M. Hashemipour: Sample collection, analysis and interpretation of data the article. All authors approved final version of the manuscript.