The complications of atherosclerosis such as coronary and cerebrovascular disease, are the most prevalent causes of mortality and morbidity worldwide. A single nucleotide polymorphism (SNP) rs1883832 (-1C/T) in CD40 gene has been recently suggested to contribute to the susceptibility to atherosclerosis in Chinese population; however, previous genetic association studies yielded inconsistent results.A meta-analysis of eligible studies reporting the association between rs1883832 and atherosclerosis in Chinese population was carried out.Pooling 7 eligible case-control studies involving 2129 patients and 1895 controls demonstrated a significant association between rs1883832 and atherosclerosis under dominant model   in Chinese population with evident heterogeneity. Meta-regression analysis indicated that the heterogeneity could be completely explained by disease category. In subgroup analysis, rs1883832 conferred ORs of 2.866  and 1.680  for coronary artery disease (CAD) under co-dominant model without heterogeneity. Similar results were obtained for acute coronary syndrome (ACS) . The other genetic models including dominant, recessive and additive models, yielded consistent results without heterogeneity for CAD and ACS, respectively. However, a protective role was found for C allele in ischemic stroke (IS) under recessive model  and additive model  with reduced heterogeneity.This meta-analysis provided evidence of association of rs1883832 C allele with an overall increased risk of atherosclerosis but distinct effect of C allele on CAD (including ACS) and IS in Chinese population, respectively. Atherosclerosis, with clinical consequences including coronary artery disease (CAD) and ischemic stroke (IS), is one of the leading causes of death worldwide We undertook the present meta-analysis of relevant literature to evaluate the association of rs1883832 (-1C/T) in CD40 gene with atherosclerosis  in Chinese population, with the following objectives: A) to estimate C allele frequencies in controls; B) to assess if the effect of rs1883832 on atherosclerosis  is statistically significant in Chinese population; if so, C) to estimate the magnitude of effect of C allele and determine the most appropriate genetic model (s) for atherosclerosis  in Chinese population.We sought eligible studies which met the following criteria: (1) published studies based on case-control design assessing the association between rs1883832 (CD40 gene -1C/T) and atherosclerosis; (2) involving subjects restricted to Chinese ethnicity; (3) providing sufficient data for calculating genotypic odds ratio (OR) with corresponding 95% confidence interval (95% CI). Non-English language studies were excluded, as were review articles, case reports, editorial comments and animal studies. It inevitably introduces a source of bias when a meta-analysis incorporates the genetic association studies from the same or different research group(s) utilizing duplicate samples. Thereby, it is obligatory to exclude in advance those repetitive publications to avoid the duplicate-study effects on our meta-analysis. The recruiting periods and hospitals were checked thoroughly for all the included studies. Additionally, the corresponding authors were contacted directly via email to confirm whether or not there were duplicate results to be eliminated.MEDLINE and EMBASE were independently searched by two reviewers (Xiaochun Ma and Chi Ma) for potentially eligible studies from the earliest available date to August 2013. The search terms used either alone or in combination included “rs1883832-1C/T, CD40, TNFRSF5, atherosclerosis, coronary artery disease, acute coronary syndrome, ischemic stroke and cerebral infarction”. Reference lists from retrieved articles were also manually searched for articles meeting our criteria. Abstracts identified using our search strategies were reviewed separately by two reviewers (Xiaochun Ma and Yan Yun). The full-text articles that potentially met criteria were then reviewed in duplicate to determine the inclusion in the analysis. Then all the authors further critically reviewed the studies upon which the 2 reviewers disagreed. Either the inclusion or exclusion of a certain study was agreed upon by all the authors finally.Data extraction was performed independently by two reviewers (Xiaochun Ma and Chi Ma). In the process of data abstraction, the two authors were blinded to the information of each research group conducting the study. Extracted data included: (1) author’s name, year of publication and journal; and (2) study design, sample size, characteristics of cases and controls , genotyping method, distribution of genotypes and alleles in cases and controls. The quality of studies was assessed with the Newcastle-Ottawa scale (NOS) for observational studies 2 test (P>0.05) only in control groups. Pooled frequency of the A allele was estimated using the inverse variance method previously described by Thakkinstian et al. 2 metric for heterogeneity were also estimated separately for crude OR1, OR2 and OR3. Three modified ORs and various genetic models would be measured afterwards only if there was no evident heterogeneity. For the Q statistic, a p value of less than 0.1 was considered statistically significant. For the I2 metric, I2>50% indicates significant heterogeneity Hardy-Weinberg equilibrium was assessed again for each study by goodness-of-fit xIf OR1 = OR3≠1 and OR2 = 1, then a recessive model is suggested.If OR1 = OR2≠1 and OR3 = 1, then a dominant model is suggested.If OR1>OR2>1 and OR1>OR3>1 (or OR1<OR2<1 and OR1<OR3<1), then a co-dominant model is suggested.Besides the finest genetic model, several other possible genetic models were also assessed: co-dominant model (modified OR1 and OR2), (C/C versus T/T) and (C/T versus T/T); dominant model (OR4), (C/C and C/T) versus T/T; recessive model, C/C versus (C/T and T/T) (OR5); additive model (OR6) (table S1). For additive model, scores of 0, 1, and 2 were assigned to genotype CC, CT, and TT respectively and per-allele ORs were calculated by logistic regression model. Sensitivity analysis was performed to assess the effects of each study , on the pooled results. Publication bias was assessed using the Funnel plot method with Egger’s test. Random-effects models were chosen for pooling data and a p value of less than 0.05 was considered statistically significant for all estimates except for Q tests of heterogeneity. Agreement assessment between reviewers was evaluated using the Cohen k statistic. Stata 12.0 and Reviewer Manager 5.1 were utilized for performing the meta-analysis.A total of 156 potential eligible citations were identified with the literature search, among which there were no articles written and published in languages other than English and Chinese. The titles of these citations were reviewed and 125 were rejected leaving 31 potentially eligible studies. After review of abstracts, an additional 14 were rejected. Seventeen studies were then retrieved in full-text to determine the inclusion. Assessment agreement of study selection between the two reviewers led to a k score of 0.84 . Then a further discussion was conducted among all the authors with regard to the studies upon which the 2 reviewers disagreed. Finally, corporate critical review of full-text excluded 4 studies for not reporting Chinese population, 3 studies for being reviews and editorial comments, 3 studies for without data for rs1883832 . All theUltimately, 7 studies (5 on coronary artery disease [CAD] and 2 on ischemic stroke [IS]) involving a total of 4024 subjects (2129 patients and 1895 controls) met our criteria and were included in this meta-analytical study 2 = 91.5%) (2 = 39.7%) (2 = 98.4%)  under random-effects model. Heterogeneity was statistically significant  . When po= 39.7%)  while th= 98.4%) .2 metric indicated heterogeneity for OR1 and OR3 but not for OR2. To explore potential sources of between-study heterogeneity, a meta-regression model was employed using the data for crude OR1. After initially running an empty regression 2 (tau2 = 0.5276), a univariate model was then performed by adding disease category (CAD and IS), which reduced the tau2 value to 0 . It indicated that the heterogeneity could be totally explained by category of atherosclerotic diseases. Pooling all the 7 studies yielded crude OR1 of 1.926 , OR2 of 1.502  and OR3 of 1.214 . Logistic regression showed that the overall gene effect was statistically significant, with the modified OR1, OR2, and OR3 being 1.896 , 1.474 , and 1.255 , respectively. Those results suggested a dominant model of gene effect on atherosclerosis   were pooled for three crude ORs and Cochran’s Q statistic and the I 2.260]) . Several 2.260]) . SubgrouFor CAD, in subgroup analysis, pooling the 5 studies  yielded crude OR1 of 2.869 , OR2 of 1.691  and OR3 of 1.688  without heterogeneity. Logistic regression showed that the overall gene effect was statistically significant, with the modified OR1, OR2, and OR3 being 2.866 , 1.680 , and 1.707 , respectively. Those results suggested a co-dominant model of gene effect on CAD . SeveralFor IS, in subgroup analysis, pooling the 2 studies  demonstrated crude OR1 of 0.684 , OR2 of 1.216  and OR3 of 0.556  with marginal heterogeneity for OR2, evident heterogeneity for OR3 and no heterogeneity for OR1. Logistic regression showed that the overall gene effect was statistically significant, with the modified OR1, OR2, and OR3 being 0.666 , 1.190 , and 0.562 , respectively. Those results suggested a recessive protective model of gene effect on IS  . SeveralFor ACS, 4 studies  were pooled to yield crude OR1 of 3.679 , OR2 of 1.983  and OR3 of 1.852  without heterogeneity. Logistic regression showed that the overall gene effect was statistically significant, with the modified OR1, OR2, and OR3 being 3.674 , 1.981 , and 1.854 , respectively. Those results suggested a co-dominant model of gene effect on ACS . ResultsSensitivity study was conducted to assess the effects of each study on the pooled results of dominant model for atherosclerosis, co-dominant model for CAD and recessive model for IS, respectively. As showed in The funnel plots and the Egger’s tests demonstrated that there was no publication bias in the dominant model for atherosclerosis  and co-dominant model and additive model for CAD  . AttemptIn the current meta-analysis, pooling the data from 7 published case-control studies indicated a significant association between a single nucleotide polymorphism (SNP) rs1883832 in CD40 gene and atherosclerosis under dominant model in Chinese population. Subgroup analysis presented that C allele of rs1883832 conferred a higher risk for CAD or ACS under all genetic models, with co-dominant model potentially being the most appropriate model. However, a protective role of gene effect was noted for C allele in ischemic stroke (IS) under recessive and additive models. Our results demonstrated that rs1883832 contributed to an overall increased risk of atherosclerosis and provided clear evidence of distinct effect of rs1883832 on CAD (ACS) and IS in Chinese population, respectively. To our knowledge, this study was the first performed to pool published case-control studies to obtain estimates for association of rs1883832 with atherosclerosis in populations of Chinese ethnicity.Genetic association studies with case-control design usually have a lack of power to replicate the associations from genome-wide association studies (GWAS). The difficulty of replication occurs partially due to the modest effect of those SNPs in complex diseases. Additionally, the overestimation of genetic effect may render replication studies underpowered and possibly fail because the necessary sample size is underestimated based on the original data from GWAS, a phenomenon known as the “winner’s curse” CD40-CD40L interaction has recently been linked with the pathogenesis of atherosclerosis. Previous studies demonstrated the elevation of serum-soluble CD40L levels which may serve as one key indicator of an independent, increased risk of major adverse cardiovascular events in patients with unstable CAD. The expression of the CD40-CD40L system on platelets and B lymphocytes was as well found to be up-regulated in patients with ACS We wish to point out a few limitations of the present meta-analysis that readers should consider when interpreting the study results. First, inclusion of retrospective observational studies based on case-control design inevitably introduced a source of potential bias that inherently exists in non-randomized, unblinded design. Second, we failed to eliminate the heterogeneity in the subgroup analysis for IS. Third, atherosclerosis is a complex disorder triggered by multiple susceptibility genes and environmental factors. However, lacking of further evaluation of gene-gene and gene-environment interactions may have a diluting influence on the estimates for rs1883832. Fourth, other SNPs within or near CD40 gene, including rs1535045, rs3765459, rs4810485, rs3092952 and rs3092920, could be constructed statistically with rs1883832 to form multi-SNP haplotypes, which fine-map this region and are potentially more informative than individual SNP. Linkage disequilibrium (LD) pattern of this region should as well be determined using genotyping data of these SNPs Our meta-analysis suggests that C allele rs1883832 in CD40 gene is positively associated with susceptibility to atherosclerosis, CAD and ACS whereas C allele appears to contribute to a decreased risk of IS in Chinese population. In the future, large scale case-control studies with rigorous design are called for in different ethnicities to replicate the association between rs1883832 and atherosclerosis and various atherosclerotic diseases. Further efforts for fine-mapping of CD40 gene region and functional analysis are required to identify causal variants and elucidate detailed mechanisms of CD40 polymorphisms in atherosclerosis.Table S1Genetic models tested in this meta-analysis. Abbreviation: OR, odds ratio.(DOCX)Click here for additional data file.Checklist S1PRISMA checklist for this meta-analysis.(DOC)Click here for additional data file.


CDH1), with conflicting results reported.Gastric cancer (GC) is a common cause of cancer-related death. The etiology and pathogenesis of GC remain unclear, with genetic and epigenetic factors playing an important role. Previous studies investigated the association of GC with many genetic variants in and promoter hypermethylation of E-cadherin gene  and other less-studied genetic variants,To clarify this inconsistency, we conducted updated meta-analyses to assess the association of genetic variants in and the promoter hypermethylation of Data sources were PubMed, Cochrane Library, Google Scholar, Web of Knowledge, and HuGE, a navigator for human genome epidemiology.CDH1 in participants with and without GC (or providing odds ratios [OR] and their variances).Study eligibility criteria and participant details are as follows: studies were conducted on human subjects; outcomes of interest include GC; report of genotype data of individual genetic variants in (or methylation status of) I2 for the assessment of between-study heterogeneity, and publication bias was assessed using funnel plot and Egger test.Study appraisal and synthesis methods included the use of OR as a measure of the association, calculated from random effects models in meta-analyses. We used P = 0.215), assuming an additive model . C-160A was associated with cardia , intestinal , and diffuse GC . The association of C-160A with noncardia GC is of bottom line significance . Multiple other less-studied genetic variants in CDH1 also exhibited association with GC. Gene-based analysis indicated a significant cumulative association of genetic variants in CDH1 with GC . Sensitivity analysis excluding studies not meeting Hardy–Weinberg equilibrium (HWE) yielded similar results. Analysis by ethnic groups revealed significant association of C-160A with cardia GC in both Asian and whites, significant association with noncardia GC only in Asians, and no significant association with intestinal GC in both ethnic groups. There was significant association of C160-A with diffuse GC in Asians (P = 0.011) but not in whites (P = 0.081). However, after excluding studies that violate HWE, this observed association is no longer significant (P = 0.126). We observed strong association of promoter hypermethylation of CDH1 with GC , suggesting that epigenetic regulation of CDH1 could play a critical role in the etiology of GC.A total of 33 studies from 30 published articles met the eligibility criteria and were included in our analyses. We found no association between C-160A and GC (OR = 0.88; 95% confidence interval [CI], 0.71–1.08; Limitations of this study are as follows: we could not adjust for confounding factors; some meta-analyses were based on a small number of studies; sensitivity analysis was limited due to unavailability of data; we could not test publication bias for some meta-analyses due to small number of included studies.CDH1 are needed to validate the results found in this study and to explore additional epigenetic loci that affect GC risk.We found no significant association of the widely studied genetic variant C-160A, but identified some other genetic variants showing significant association with GC. Future studies with large sample sizes that control for confounding risk factors and/or intensively interrogate CpG sites in  Based on anatomic conditions, GC can also be divided into 2 subtypes: gastric cardia cancer and noncardia GC, with the former referring to cancers of the top portion of the stomach and the latter referring to cancers in the other areas of the stomach. Noncardia cancer is commonly associated with the Helicobacter pylori infection. There was no overall association between gastric cardia cancer and H pylori infection, whereas a positive association was observed in high-risk populations.4Gastric cancer (GC) is one of the most common gastrointestinal malignancies throughout the world. Over the past half century, the incidence of GC has gradually decreased. However, GC remains to be the second most common cause of cancer-related death, with >700,000 deaths/y.H pylori infection,5 inadequate vitamin C uptake,6 smoking,7 high salt intake,8 and low vegetable intake.9 Meanwhile, multiple genetic variants and different genetic pathways have been identified to contribute to GC risk,10 suggesting that genetic factors play important roles in GC susceptibility. Many studies have been conducted to search for susceptibility genes for GC, such as Interleukin-1, Interleukin-8, Glutathione S-Transferase, and Cytochrome P450 2E1.11The etiology and pathophysiology of GC is not fully understood. It is well established that gastric carcinogenesis is a complex multifactorial and multistage process. Previous studies have identified several risk factors that might contribute to gastric carcinogenesis including CDH1), is involved in the establishment and maintenance of intercellular adhesion.12 In vitro studies found that the A allele of C-160A could decrease the transcriptional efficiency of CDH1 by approximately 70%, suggesting that the A allele could potentially increase susceptibility to GC.13 Many previous studies investigated the association of the genetic variants, C-160A (rs16260) in CDH1 with GC risk, with conflicting results reported. Several meta-analyses have also been conducted to examine the association of C-160A with GC. Although all of them found no significant association of C-160A with GC, subgroup analysis by ethnic groups reported inconsistent findings  may only confer a small or marginal individual effect on GC susceptibility. Studies focused on individual genetic variant may be less powerful in detecting small genetic effect and fail to capture the joint contribution from multiple genetic variants. We therefore conducted a gene-based analysis to examine the cumulative effect of multiple genetic variants in CDH1 on GC risk.E-cadherin glycoprotein, encoded by E-cadherin gene CDH, is invoCDH1. Details of keywords used in the literature search can be found in the supplementary file . We used the following inclusion criteria in determining study eligibility: studies on human subjects, outcomes of interest include GC, and report of genotype data of individual genetic variants in (or methylation status of) CDH1 in participants with and without GC (or providing odds ratios [ORs] and their variances). All potentially relevant publications were retrieved and further evaluated for inclusion. We also hand-searched references of all relevant publications for additional studies missed by the database search. Only studies published in the English language were included in our analysis. Two authors (H.J. and J.Y.Y.) performed the search independently. Disagreement over eligibility of a study was resolved by discussion until a consensus was reached.From January to May 2014, we did an extensive literature search in PubMed, Cochrane Library, Google Scholar, Web of Knowledge, and HuGE, a navigator for human genome epidemiology, for candidate gene studies on the association of GC with genetic variants in and promoter hypermethylation of Two reviewers (J.Y. and L.M.) independently extracted the following data according to a prespecified protocol: first author’s name, year of publication, characteristics of the study participants , genotype or methylation status data for subjects with and without GC (or OR and the corresponding variances), and the genetic model used . Discrepancies were resolved by discussion, and extracted data were entered into a computerized spreadsheet for analysis.CDH1 and GC. We used random effects models to calculate ORs and the corresponding 95% confidence intervals (CIs). The inverse of the variance of each study was used as the weight for that study. We used forest plots to graphically represent the calculated pooled ORs and their 95% CIs. Each study was represented by a square in the plot, the area of which is proportional to the weight of the study. The overall effect from the meta-analysis is represented by a diamond, with its width representing the 95% CI for the estimate. We used I2 for assessment of between-study heterogeneity, and publication bias was assessed using funnel plot and Egger test, and a P value <0.20 was considered statistically significant.We used the OR as a measure of the association between the genetic variants in and methylation status of CDH1 with GC, when there are multiple eligible studies for the genetic variants. Otherwise, we compiled the results of the association with GC for genetic variants that appear in single studies. We also analyzed the association between C-160A and subtypes of GC . Meta-analyses were conducted when there were multiple studies for the analysis of each subtype.We performed an updated meta-analysis for the association of C-160A with GC, and also conducted meta-analysis for association of other genetic variants in CDH1 with GC, we conducted a gene-based analysis using the P values for the association of individual genetic variants in CDH1 with GC, calculated from our meta-analyses and/or from published literature. We used 4 popular P value combination methods to assess this cumulative association: the Fisher method,23 the Simes method,24 the modified inverse normal method,25 and the truncated product method (TPM).27 A detailed description of the 4 methods has been reported elsewhere.28 We used 100,000 simulations to estimate the combined P value for TPM because the individual P values are most likely to be dependent.In order to assess the cumulative association of CDH1 on susceptibility of GC.Finally, we performed meta-analysis to examine the effect of promoter hypermethylation of We performed separate meta-analyses by excluding studies in which genotype in the control group did not meet Hardy–Weinberg equilibrium (HWE). We also performed meta-analysis separately for individual ethnic groups/countries of origin (Asian and whites).29 Meta-analysis was performed using Stata 11.2 . All other analyses were performed using SAS version 9.3 , R (www.R-project.org), and Matlab 8.1.0.604 . A P value <0.05 was considered statistically significant.As a research using systematic review and meta-analysis, ethical approval of this study is not required. This work was reported according to the PRISMA guidelines.57Figure CDH1 was reported in individual studies . These results, together with results obtained from our mea-analyses, were included in our gene-based analysis. Moreover, the association of the promoter hypermethylation of CDH1 with GC has been examined in 8 studies, and meta-analysis was performed to explore the effect of promoter hypermethylation on GC risk. DNA methylation was measured similarly across studies .All qualified publications were published since 2002 and had sample sizes ranging from 14 to 1197 . There was some evidence of publication bias for the meta-analysis for promoter hypermethylation of CDH1 . Due to space limitations, we only present the results using an additive model. Results obtained using other models can be found in the supplementary materials.We calculated the association between the C-160A in 33 indicated that compared with CC carriers, those carrying each additional copy of the A allele had increased risk of GC, whereas the other 7 studies reported decreased risk. Our meta-analysis indicates no significant association of C-160A with GC .Of the 22 studies included in our meta-analysis, 10 showed significant association between C-160A and GC Table . Specifi−6), intestinal , and diffuse GCs . The association of C-160A with noncardia GC is of bottom line significance  from another individual study46 also exhibited significant association with GC (P = 7.53 × 10−5).Our meta-analysis of the less-studied genetic variants in GC Table . HoweverCDH1 with GC, we performed a gene-based analysis using all the P values we obtained for each individual genetic variant in CDH1. Additionally, we examined whether the association varies in meta-studies only  and in individual studies only (including only results for genetic variants that appeared in single studies). Our gene-based analysis indicated a significant association between the genetic variants in CDH1 and GC . The association held when pooling results from only individual-studies, but disappeared when only results from meta-studies were included, indicating that the observed gene-based association was driven mainly by results from the less-studied genetic variants . We would like to caution against over interpretation of the results from individual studies because, due to inadequate number of studies, it is not possible to determine whether there is selective reporting that can lead to inflation of the P values.To examine the cumulative association of multiple genetic variants in CDH1 with GC , and another study48 showed marginal association (P = 0.080). All the other 6 studies indicated that promoter hypermethylation of CDH1 is significantly associated with increased risk of GC.Our meta-analysis of 8 studies showed very strong and significant association of promoter hypermethylation of 45 After excluding these 4 studies, our meta-analysis again indicated no significant association of C-160A with GC . There were 12 studies from 11 articles and 8 studies from 6 articles examining the association of C-160A with GC in Asian and white participants, respectively. Our meta-analysis based on these studies found no significant association of C-160A with GC in Asian  and white participants . These results do not change after excluding studies that did not meet HWE.There were 4 studies in which the genotype in the control group did not meet HWE.30 and 2 studies41 examining the association of C-160A with cardia GC in Asian and white participants, respectively. A significant association was found between C-160A and GC in both Asian  and white participants . There were 2 studies examining the association of C-160A with noncardia GC in Asian35 and white participants.42 Our meta-analysis found significant association in Asian  but not in white participants . There were 3 studies44 and 2 studies41 examining the association of C-160A with intestinal GC in Asian and white participants, respectively. Our meta-analysis based on these studies found no significant association in Asian  and white participants . There were 6 studies and 3 studies examining the association of C-160A with diffuse GC in Asian and white participants, respectively. Our meta-analysis based on these studies indicated significant association of C-160A with diffuse GC in Asian participants  but not in white participants . However, after excluding the studies in which genotype in the control group did not meet HWE (n = 2), the association in Asian participants is no longer statistically significant .There were 1 studyCDH1. We provided an updated meta-analysis on the widely studied genetic variant C-160A. Our analysis showed that C-160A is not associated with GC, either in the overall population, or in Asian or white participants. However, within a very limited set of articles that evaluated subtypes of GC, we found significant association of C-160A with cardiac, intestinal, and diffuse GC. We found that the promoter hypermethylation of CDH1 is strongly associated with GC, indicating potential epigenetic influences in the carcinogenesis and development of GC. To the best of our knowledge, this is the most comprehensive meta-analysis on the association of GC with a number of genetic variants in CDH1, and with promoter methylation of CDH1.In this study, we conducted an extensive literature search for publications on the association of GC with genetic variants in and promoter hypermethylation of I2 = 93.0%; 95% CI, 90.6%–94.7%). Identifying the source of heterogeneity is challenging with limited information provided in many studies. Variation in patient characteristics might be an important source of heterogeneity. Some studies used matched controls ,46 whereas most other studies did not perform matching. Other patient characteristics, such as smoking behavior, H pylori infection, and tumor location, can also contribute to the heterogeneity of the included studies in the meta-analyses.In the meta-analysis of C-160A, we identified significant heterogeneity between the studies included for analysis . Realizing that the observed gene-based association might be driven by the results reported in the study by Jenab et al,38 in sensitivity analysis we dropped that study from the gene-based analysis and still observed significant gene-based association . The SNP rs1125557 is in high linkage disequilibrium (LD) with C-160A . Given the high LD, we feel that the significant finding was probably because of the small sample size based on which the result was reported.46 Studies on functional outcomes of these less-studied genetic variants in CDH1 are scarce, and further studies are needed to elucidate whether and how they function in influencing disease susceptibility.Of the 24 less-studied genetic variants in 58 The precise molecular mechanism underlying the association of promoter hypermethylation of CDH1 with GC remains to be understood. A key challenge remains whether changes in methylation are a cause or an effect of the pathological process. Although some studies suggest that altered methylation in CDH1 might be involved in carcinogenesis of GC but not development of GC,55 others indicate that accumulation of aberrant methylation might be an important mechanism for GC development.48 There are also studies indicating that the accumulation of DNA methylation might be caused by proliferative changes during tumor progression.49 Moreover, CDH1 methylation seems to be age related,51 making it more complicated to disentangle the exact role of methylation in the initiation and development of GC. More future large-scale studies are needed that examine subjects at risk of developing GC as well as subjects with GC to better elucidate whether and how CDH1 promoter hypermethylation is implicated in GC initiation and development.DNA methylation is the most extensively studied epigenetic modification, and plays an important role in regulating gene expression and cell differentiation. Aberrant DNA methylation leads to silencing of tumor suppressor genes or loss of oncogene repression, and therefore is an important mechanism in the initiation and development of GC.H pylori infection. First, future studies are needed to validate our results—especially large consortium studies that provide control for such confounding factors. Second, some meta-analyses were based on few studies, and the gene-based analysis used some results from individual studies. Third, sensitivity analyses by ethnicity are limited because race information was not available in all studies. Fourth, due to the limited number of studies included in some of the meta-analyses, we could not test publication bias for them. This might lead to bias in the resulting data, and subsequently influence the validity of the gene-based analysis. Finally, there are other types of genetic variations that are not included in our study, such as copy number variation that was recently reported to be associated with GC.60Our study has some limitations. Since relevant data were not available, our meta-analysis could not adjust for confounding factors such as age, sex, smoking behavior, or CDH1 on GC risk. We found no significant association of the widely studied genetic variant C-160A with GC. However, a limited number of studies suggest that C-160A may be associated with subtypes of GC in different ethnic groups, and we identified some other genetic variants showing significant association with GC. Gene-based analysis indicated that the previously studied variants cumulatively influence GC susceptibility. Meta-analysis on the promoter hypermethylation of CDH1 suggests that epigenetics also plays a critical role in the carcinogenesis of GC. Future studies with large sample sizes that control confounding risk factors and/or intensively interrogate CpG sites in CDH1 are needed to validate the results found in this study and explore additional epigenetic loci that affect GC risk.In summary, in this study, we performed meta-analyses to analyze the genetic and epigenetic effect of


Background: Migraine is a common neurovascular disorder with multifactorial and polygenic inheritance. The aim of this study was to investigate the association of a migraine without aura and Ala379Val polymorphism of lipoprotein-associated phospholipase A2 (Lp-PLA2) gene in the Iranian population.Methods: In this study, 103 migraine patients and 100 healthy controls were enrolled. DNA samples were extracted and the Ala379Val polymorphism of Lp-PLA2 gene was investigated. To assess severity of a headache, patients filled out the headache impact test (HIT-6) and migraine severity (MIGSEV) questionnaires.Results: Allele V had significantly lower frequency in the case group than control subjects . The frequency of migraine patients that were a carrier of V allele (V/V and A/V) was statistically significant lower than the control group . There was no significant difference of alleles frequency between three grades of MIGSEV (P = 0.316). Furthermore, total HIT-6 score was not significantly different between different genotypes (P = 0.466).Conclusion: Our results showed that Ala379Val gene polymorphism of LP-PLA2 is associated with lower risk of migraine but not with severity of headaches in an Iranian population. It affects about 12% of the adult population which is more prevalent in females.6,7 It is known that mass and activity of Lp-PLA2 associated with race (white is more than black) and sex .8 Lp-PLA2 is an enzyme that circulates in blood and classified as a new inflammatory marker and atherosclerosis risk factor because of the production of oxidized free fatty acids and lysophosphatidylcholines during oxidation of low-density lipoprotein.6,9 The proatherogenic effect of Lp-PLA2 may be through its role in the generation of lysophosphatidylcholine and oxidized fatty acids and also hydrolysis of the proinflammatory mediator PAF.10 Lp-PLA2 activity and mass were correlated with increased ultrasound determined carotid intima-media thickness. Moreover, there is an association between Lp-PLA2 and cardiovascular diseases.7,8Lipoprotein-associated phospholipase A2 (Lp-PLA2), also known as platelet-activating factor (PAF) acetylhydrolase, is a calcium-independent serine lipase involved in so many physiological actions and is produced by inflammatory cells such as macrophages, monocytes, and T-lymphocytes.11,12 The different polymorphisms of Lp-PLA2 have association with some diseases. For example, G994T has an association with polycystic ovary syndrome and also has increased susceptibility to oxidative stress and inflammation.13The Lp-PLA2 gene is located in the 6 q21.2-p12 chromosome and different polymorphisms have been described for this gene. Studies have shown that some polymorphisms exist in one race that is not shown in others. Val279Phe and Gln281Arg are two polymorphisms of this gene that have only seen in Japanese populations; however, Arg92His, Ile198Thr and Ala379Val, other polymorphisms of this gene have been described more in European populations.14 By reducing the Lp-PLA2 protein production, the risk of coronary heart disease (CHD) may decrease. Studies have shown that in individuals homozygous for Ala379Val this enzyme production had reduced and it may lead to less risk for CHD.15 As migraine is a vascular disorder and Lp-PLA2 has some vascular effects, the aim of this study was to investigate the association of migraine without aura and Ala379Val polymorphism in Iranian population.Studies have shown that V279F variant in the Lp-PLA2 gene is associated with the increase of myocardial/cerebral infection and also shown an association between this variant and an increased of coronary artery disease (CAD) and infarction. On the other hand, some other polymorphism of this gene like V379 allele of the Ala379Val polymorphism reduced the risk of CAD.16 and also controls that were matched for age, education, sex, and socio-economic status with the cases. The patients were selected from three outpatient neurology clinics between November 2011 and June 2012 in Isfahan, Iran. The socio-demographic and headache characteristics of all the subjects including age, sex, level of education (with and without academic degrees), residency , frequency of headaches, the effect of menstruation on the migraine, and positive family history were asked. Patients who had at least a 3-month history of headaches before the diagnoses and had not have received any medications for their migraine were consecutively enrolled. Controls were selected from healthy people who accompanied the patients that were referred to the neurologic clinics (including patients with migraine and other neurologic disorders) who did not have any history for migraines and also any family history of migraine in the first degree relatives. An informed consent was taken from participants before entering the study. We conducted a case-control study enrolling patients that were recently diagnosed as without aura migraines based on the International Headache Society (IHS) criteriaHeadache impact test (HIT-6) questionnaire17To assess the severity of headaches, each patient filled HIT-6 questionnaire. HIT-6 questionnaire consists of six questions that show six dimensions: pain severity, role functioning, social functioning, vitality, emotional distress, and cognitive functioning. Each question has 5 available answers: never, rarely, sometimes, very often, and always. Choices have been scored between 6 and 13. The total score is between 36 and 78.Migraine severity (MIGSEV) questionnaire18 is a simple severity scale with four items including intensity of pain, disability in daily activity, tolerability and nausea that categorize patients in three groups of intensity: mild, moderate, and severe. This instrument is highly reliable, reproducible and sensitive, and we have used its Persian translation in our previous study as a valid scale.18,19Each migraine patient completed an MIGSEV questionnaire as a valid scale for assessing headache severity. The MIGSEV scale, developed by EL Hasnaoui in 2003,DNA extraction and genotypingA value of 2 ml of venous blood was collected from each participant. Genomic DNA samples were extracted from peripheral whole blood using the AccuPrep Genomic DNA Extraction kit  according to the manufacturer’s protocol.The single-nucleotide polymorphisms (SNPs) rs1051931 (A379V) were identified by the National Center for Biotechnology Information (NCBI) data bank and primers were designed by Beacon Designer 8.00 to flank the coding regions . The forward primer was 5’-TTCTCTTTTAGGGGTTCAGT-3’ and reverse primer was 5’-CATCTGGTTTAGGTCATGAAAA-3’. Genotyping was done by high-resolution melt (HRM) assay using a Rotor-Gene 6000 instrument . Polymerase chain reactions (PCRs) were carried out in duplicate in 20 µl of final volume using the type-it HRM kit (Qiagen), HRM PCR buffer, HotStarTaq Plus DNA Polymerase, nucleotides and EvaGreen dye, and 30 ng DNA. The PCR program consisted of an initial denaturation-activation step at 95 °C for 5 minutes, followed by a 40-cycle program . Curves for each duplicate were checked on the shape and peak height to meet reproducibility. Normalized and temperature-shifted melting curves from HRM, suggestive of SNPs, were distinguished and the samples were subjected to direct sequencing.We analyzed our data with SPSS software . An independent t-test was used for quantitative variables between two groups. The relation between polymorphism (homozygous and heterozygous) and different categorized variables,  was established using chi-square test and calculation of odds ratio (OR) [confidence interval (CI) = 95%]. The significant level was considered as P < 0.050.The study was approved by the Ethical Committee of Isfahan University of Medical Sciences.In this study, 103 subjects with migraine and 100 healthy subjects were enrolled, and DNA samples were analyzed for LP-PLA2 Ala379Val gene polymorphisms. In the case and control group, 82.6 and 78.0% were female, respectively, and distribution was not statistically significant. Total mean age of the subjects was 34.42 ± 0.73 . There were no significant differences in the mean ages, education level (40.0 vs. 36.0% without academic degrees) and residency (51.5 vs. 57.6% urban) between case and control groups, respectively. The migraine characteristics of the case group are shown in The analysis of allele frequency in the case group shows no significant difference in the distribution of A and V alleles in the two genders (P = 0.282), but in control group, V allele frequency in male subjects was significantly more than females (P = 0.008). Another analysis in the case group showed that the distribution of alleles A and V was not associated with family history of migraine and also menstruation effect . The patients were classified into three groups according to the MIGSEV grade. There was no significant difference of alleles frequency between three grades of MIGSEV (P = 0.316).The total frequency of allele A and V in the study population were 287 and 119, respectively. Allele V had lower frequency in the case group than control subjects, and the difference was statistically significant  . The freThe frequency of migraine patients that were carrier of V allele was statistically significant lower than control group . 20As migraine is a vascular disorder and Lp-PLA2 has some vascular effects, the aim of this study was to investigate the association of migraine without aura and Ala379Val polymorphism in Iranian population. Our results showed that frequency of V allele  was lower in case group than control group. We classified our subjects into three genotypes of A/A, A/V, and V/V. The frequency of migraine in the subjects who are carrier of V allele was statistically significant lower than non-V allele carriers , so the current study suggests that V allele of Ala379Val polymorphism of LP-PLA2 will be related to decreased risk for migraine. We also investigated the association of Ala379Val polymorphism with severity and frequency of headaches in migraine patients. Our results revealed no association between the polymorphism and mentioned factors. To the best of our knowledge, there is no study conducted to assess the association of Ala379Val polymorphism of LP-PLA2 and migraine. But according to results of a study conducted on the genetic of migraine although they could not find a relation between Lp-PLA2 gene and migraine, but they have reported that a predisposing haplotype spanning 10 Mb on the chromosome 6p12.2-p21.1 that contains the Lp-PLA2 gene, was inherited with all migraine patients in the pedigree that they have studied.15 Similar to this report, our results showed that there was no significant difference between AA and AV groups in the frequency of migraine. However, VV genotype was associated with lower risk of migraine. Results of another study conducted by Oei et al. revealed that activity of Lp-PLA2 is associated with risk of CHD and also ischemic stroke.21 This report confirms the Abuzeid et al.15 conclusion as we know that V379 allele is associated with lower activity of Lp-PLA2 protein.21There are some other studies designed to assess association of other disorders and Ala379Val polymorphism of LP-PLA2. Abuzeid et al. in their study have reported that homozygosity for the V379 allele was associated with lower risk of myocardial infarction (MI), but the risk of MI in the AA and AV subjects was very similar.22 that was designed to investigate the association of Ala379Val polymorphism with hypertension and thrombotic markers, results showed that A/V genotype is associated with lower risk of essential hypertension. But both A/A and V/V genotypes were at higher risk of developing essential hypertension. Their results also showed that in hypertensive patients, A/V genotype carriers have elevated level of fibrinogen. They have reported that in hypertensive patients, A/V genotype is associated with increased inﬂammatory and thrombotic burden.In a study by Kardara et al.Our results showed that Ala379Val gene polymorphism of LP-PLA2 is associated with lower risk of migraine without aura in Iranian population. However, it has no effect on the severity of migraine and also frequency of headaches in migraine patients. We had a small sample size and it reduced the power of our study especially in analysis of association of the polymorphism with frequency of headache and severity of migraine. Therefore, this is a preliminary conclusion. Further studies with larger sample sizes are needed.


On the other hand, being underweight , obese , and class II/III obese  after diagnosis were associated with significantly increased all-cause mortality. Being obese prior to diagnosis of colorectal cancer was associated with increased colorectal cancer-specific mortality and all-cause mortality, whereas being obese after diagnosis was associated with increased all-cause mortality. The associations with being underweight may reflect reverse causation. Maintaining a healthy body weight should be discussed with colorectal cancer survivors.Studies have reported conflicting results on the association between body mass index (BMI) and prognosis of colorectal cancer. Therefore, we have conducted a meta-analysis of prospective studies, which examined the association of pre- and post-diagnostic BMI with colorectal cancer-specific mortality and all-cause mortality in patients with colorectal cancer. We searched Medline and EMBASE database published between 1970 and September 2014. A total of 508 articles were identified, of which 16 prospective cohort studies were included for the current meta-analysis. The analysis included 58,917 patients who were followed up over a period ranging from 4.9 to 20 years (median: 9.9 years). We found that being underweight before cancer diagnosis was associated with increased all-cause mortality (Relative risk [RR]: 1.63, 95% CI: 1.18–2.23,  Colorectal cancer has become the third most common cancer in the world, making it the fourth leading cause of cancer mortality . A recenadenomas . AnotherUnderstanding the association between BMI and the prognosis of colorectal cancer is highly important to provide body weight guidelines for colorectal cancer patients. Studies have clearly identified that being underweight is associated with increased risk of death, probably due to cancer progression-associated weight loss –10. HoweRecently, Parkin et al.  compreheThis meta-analysis study followed the guidelines provided by the preferred reporting items for systemic review and meta-analyses . We condThe search terms (used in various combinations) were ‘body mass index (BMI)’ ‘colorectal cancer’, ‘colon cancer’, ‘rectal cancer’ ‘mortality’, ‘survival’, ‘over all survival, ‘cancer-specific survival’, ‘disease free survival’, and ‘prognosis free survival’. Also, we completed a manual search of references cited in the selected articles and review articles to explore for any further relevant studies. All potentially relevant studies were archived in an Endnote X6 database.The selected articles were independently screened in an unblinded standardized manner by two authors (JL and JYJ). Any discrepancies regarding eligibility for study selection were re-evaluated by further discussion involving all four authors  in order to reach consensus. The studies were assessed for eligibility using both inclusion and exclusion criteria. The inclusion criteria required that studies had a prospective study design and contained data that addressed all-cause mortality and colorectal cancer-specific mortality, with data on pre-diagnosis BMI or post-diagnosis BMI. Two case-control studies, which were converted to a survival cohort with recalled body weight, were included in this analysis ,12. In tThis analysis followed the Meta-analysis of Observational Studies in Epidemiology (MOOSE) guidelines for meta-analysis of observational studies . To ensuThe Newcastle-Ottawa Scale (NOS) procedure was used to assess the quality of the study . The NOSI2 statistic. When performing meta-analysis, fixed effect models were used when selected studies for meta-analysis were homogenous. On the other hand, random-effect models with forest plots were used for meta-analysis when selected studies were not homogeneous. In further sub-analysis, additional meta-analyses were performed for genders and site of diseases. To assess for publication bias, a visual inspection of the funnel plot was conducted to find the relationship between the study results and precision. Trim and fill analyses were used to test the potential influence that unpublished studies could have on the summary RR estimates. Statistical significance was tested using a p-value of < 0.05. All statistical analyses were performed using Comprehensive Meta-Analysis Version 1.25 software .Individual study RRs were directly extracted from the published reports, and inserted into the Comprehensive Meta-Analysis Version 1.25 software program that computed fixed and random-effect model parameters and 95% confidence intervals (CIs). Statistical heterogeneity across the sampled studies was tested using the Q statistic, and inconsistency was quantified by the Using explicit inclusion and exclusion criteria, 16 articles were selected for this meta-analysis. Details of the selection process are presented . The ini2 , < 20 kt et al. , Sinicrot et al. , and Sint et al. ), < 21 kt et al.  and Meyet et al. ), and < o et al.  and Boyl2) and class II/III (BMI ≥ 35 kg/m2). To address these variances in categories, an additional meta-analysis was conducted to determine whether the diverse categories for the reference group influenced the results. The additional meta-analysis was conducted to compare results involving all studies without considering the differences in reference categories  and obesity categories. This comparison analysis was conducted in most cases, but not in every case because the number of studies did not allow the completion of every possible meta-analysis. In none of the comparison was there evident statistically significant heterogeneity.The ranges used to define obesity categories also varied among studies. Four studies ,12,13,23p < 0.01). Pre-diagnosis overweight was not associated with colorectal cancer-specific mortality and all-cause mortality. Pre-diagnosis obesity was significantly associated with increased colorectal cancer-specific mortality  and all-cause mortality . There was no evidence of publication bias in any analyses. Additionally, there was no apparent influence of unpublished data in any analyses using the trim and fill method.Six studies reported an association between pre-diagnosis BMI and colorectal cancer-specific mortality and all-cause mortality . Pre-diap < 0.01). Post-diagnosis overweight was associated with significantly improved colorectal cancer-specific mortality . Though post-diagnosis overweight was significantly but modestly associated with improved all-cause mortality . Since studies included in examining the association between post-diagnosis overweight and the risk of all-cause mortality were not homogeneous, we used random effect models and further analyzed the association between overweight and the risk of mortality according to different categories for normal BMI (18.5–24.9 kg/m2 vs. 20–24.9 kg/m2 vs. 21–24.9 kg/m2 vs. < 25 kg/m2). When we only included studies that used 20 or 21 kg/m2 as the lower limit for normal BMI range, the reduced risk of mortality observed among overweight participants no longer existed .Twelve prospective cohort studies were included in this analysis . Post-dip < 0.01) while no association was found between post-diagnosis obesity and colorectal cancer-specific mortality. We further analyzed the association between class II/III obesity (BMI > 35 kg/m2) and the risk of mortality and found significantly increased risk of all-cause mortality . There was neither evidence of publication bias in the analyses, nor apparent influence of unpublished data in any analyses using the trim and fill method.Post-diagnosis obesity was significantly associated with all-cause mortality  and in males . Post-diagnosis overweight was not significantly associated with increased all-cause mortality in females, but slightly associated with significantly reduced all-cause mortality in males . Post-diagnosis obesity was significantly associated with increased all-cause mortality in females  but only borderline statistically significant in males . No evidence of publication bias and no apparent influence of unpublished data using the trim and fill method was observed in any analyses.A subgroup analysis was performed to examine whether the association between post-diagnosis BMI and risk of mortality differed according to gender . Post-diagnosis overweight was not significantly associated with all-cause mortality . Post-diagnosis obesity was associated with significantly increased all-cause mortality in colon cancer . The analyses did not show evidence of publication bias nor any apparent influence of unpublished data using the trim and fill method.Because there were insufficient studies to allow subgroup analysis for rectal cancer, this analysis only included colon cancer. Post-diagnosis underweight was significantly associated with all-cause mortality in patients with colon cancer  and class II/III obesity (BMI ≥ 35 kg/m2) the increased risk of mortality.Apparently conflicting results on the association between BMI and the risk of mortality in individual studies may have resulted from the timing of BMI measurement, specifically whether it was before or after the cancer diagnosis. Therefore, we performed a meta-analysis of prospective cohort studies to distinguish the influence of pre- and post-diagnosis BMI on the risk of mortality in patients with colorectal cancer. Our analysis showed that pre-diagnosis underweight and obesity were associated with increased all-cause mortality and post-diagnosis underweight, class I obesity (BMI 30–34.9 kg/m2). It is also important to understand that one of the main causes of death in obese colorectal cancer patients is cardiovascular disease rather than cancer recurrence [We found that being obese before cancer diagnosis was associated with increased colorectal cancer-specific mortality as well as all-cause mortality and being obese post-diagnosis was associated with all-cause mortality in colorectal cancer survivors. In analyzing the impact of post-diagnosis BMI and the risk of mortality, it is important to understand the reason for the body weight change. There are two reasons for weight loss in colorectal cancer patients who completed standard adjuvant therapy: some patients intentionally lose weight through a healthier diet and exercise, whereas other patients experience weight loss due to progression of cancer or effects of treatment. Post-diagnosis BMI cannot distinguish between these reasons for weight loss, which might account for the lack of association between post-diagnosis BMI and the risk of mortality. However, in our meta-analysis of ten studies, we were able to find significant association between pre-diagnosis obesity and the risk of mortality. We further found higher relative risk of death due to all cause among patients with class II/III obesity , some studies included in our meta-analysis used different classifications for normal BMI such as 20 or 21 kg/m2 as the lower cutoff. When we excluded studies that used BMI 18.5 kg/m2 as the lower cutoff for the normal range, we observed that post-diagnosis overweight was not significantly associated with reduced risk of mortality. A possible reason is that this exclusion reduces the number of people in the reference group with pre-existing disease or who have lost weight due to progressive colorectal cancer.Our finding that being overweight was associated with a lower risk of mortality may not necessarily be causal but rather reflect that the reference group  may include people who might have lost weight due to disease. Although the normal BMI category of 18.5–24.9 kg/mp = 0.03). Since physical activity, exercise and reducing body fat have positively influence on these factors, healthier lifestyle including exercise and maintaining health body weight should be recommended to improve prognosis of colorectal cancer patients [The biological mechanism that underlies the association between obesity and colorectal cancer mortality is unclear; but several studies have addressed possible mechanisms, mainly related to obesity-related hormonal changes. Obesity is associated with elevations in insulin, free insulin-like growth factors (IGFs), and adipocyte-derived factors that include leptin, TNF-alpha, IL-6, and reductions in adiponectin ,27. Manypatients . Severalpatients ,37.There are strengths and limitations of the current study. First, we only included prospective cohort studies that would be less susceptible to selection and recall biases. In addition, we have performed meta-analysis separately addressed BMI before and after diagnosis to obtain association with all-cause and colorectal cancer-specific mortality. Combining studies allowed us greater power to observe associations.We also have several limitations. The major limitation is the possibility of reverse causation, which probably precludes us identifying a pure group of healthy normal weight persons. Thus, we likely underestimated the true impact of obesity on colorectal prognosis because progressing cancers typically cause weight loss rather than weight gain. A second limitation is that pre-diagnosis and post-diagnosis BMI are correlated with each other and therefore we cannot precisely say when of obesity is acting. Nonetheless, since pre-diagnosis BMI is a major determinant of post-diagnosis BMI, it is desirable to maintain normal weight throughout life rather than relying on weight loss after diagnosis. Another limitation of the study is that several studies in this meta-analysis used self-reported BMI, though limitation is probably minor as self-reported BMI was shown to be highly correlated with measured BMI . FinallyThere are additional considerations that might influence the association between body mass index and colorectal cancer mortality in both pre-diagnosis and post-diagnosis. These factors include disease severity at diagnosis, prognosis factors, and extend to discern how findings may differ across various population groups. The current meta-analysis used RRs that were adjusted for those factors, including disease severity at diagnosis, prognosis factors, sex, age, etc. However, the exact same factors were no uniformly applied throughout all studies. Future meta-analysis studies should conduct more stratified analyses that examine a variety of populations and examine the effects of other adjustment factors. Additionally, B-Catenin status, which plays an important role in carcinogenesis of colorectal cancer, also needs to be studied to better understand the mechanism related to adiposity, physical activity, and colorectal cancer ,39.In conclusion, the findings of this meta-analysis suggest that both pre- and post-diagnosis underweight and obesity are associated with increased risk of mortality. Maintaining a normal body weight should be considered by all individuals including colorectal cancer patients. Intervention studies on the impact of weight control on the risk of mortality in colorectal cancer patients are needed.S1 TableThis study used PRISMA guidance to help improve reporting quality of this meta-analysis study.(PDF)Click here for additional data file.


Recent research displays that breast cancer (BC) is a heterogeneous disease and distinct molecular subtypes yield different prognostic outcomes.We conducted a meta-analysis to clarify the role of molecular subtypes in recurrence risk after breast-conserving therapy (BCT). Eligible studies of single-  and triple-molecular  subtypes were identified through multiple search strategies. Pooled hazard ratios with 95% confidence intervals were calculated to assess this research topic.Fifteen studies involving 21,645 participants were included in the meta-analysis. Her-2 positive patients had a significantly higher recurrence risk in both overall merge  and subtotal merge of local recurrence (LR) . Significantly higher risk of recurrence was also observed in p53 positive patients by overall merge  and subtotal merge of LR . When setting Luminal A as a baseline, Luminal B, Her-2, and triple-negative all showed significantly increased risk for both LR and distant recurrence (DR). Comparing triple-negative and non-triple-negative subtypes showed the biggest risk for overall recurrence  and LR .Our meta-analysis showed significant differences in recurrence risk among various molecular subtypes after BCT. Although Her-2 and p53 positive subtypes can be considered independent prognostic biomarkers for indicating high LR risk, triple-molecular biomarkers showed higher clinical value. Triple-negative subtype showed the highest recurrence risk among all subtypes, and adjuvant chemotherapy should be considered for it. Breast-conserving therapy (BCT) is considered the standard treatment for early-stage breast cancer (BC)-3. ThougThe factors affecting recurrence are complex, including clinical and histological characteristics, with or without postoperative adjuvant radiotherapy and systemic therapy (chemotherapy and/or hormone therapy). Recent Original articles analyzing the hazard ratio (HR) of recurrence after BCT in different BC molecular subtypes were searched by online databases PubMed, Embase, and Web of Science. We selected studies carefully by the following sets of key words variably combined: ‘breast cancer’ , ‘breast-conserving surgery’ , ‘breast-conserving therapy’ , ‘recurrence’ , ‘hazard ratio’ , ‘molecular marker’ , and ‘molecular subtype’; the last search update was performed on 10 January 2014. All eligible studies published in English were reviewed, and their bibliographies were also examined for other relevant publications. Relevant review articles were manually searched to find additional eligible studies. If more than one article was published using the same series of study subjects, we only chose the latest or most complete study for this meta-analysis. All the studies enrolled in this meta-analysis have been performed with the approval of an appropriate ethics committee. Researches carried out on humans are all in compliance with the Helsinki Declaration.We followed the guidelines of the critical checklist proposed by the Dutch Cochrane Centre Meta-analysis of Observational Studies in Epidemiology (MOOSE). ArticleAll data were carefully extracted from eligible publications in duplicate by two co-authors (PJ and JZ). Any disagreements were resolved by discussion between the two authors. The extracted data elements in Table The extracted data elements in Table 2) was used to quantify the extent of heterogeneity explained by these characteristics of enrolled studies. If significant heterogeneity was observed (P <0.1 or I2 >50%), a random-effects model (Der Simonian- Laird method) was applied; otherwise, the fixed-effects model (Mantel- Haenszel method) was adopted[All statistical analyses were conducted using Stata  and Excel . The aggregation of HRs and 95% CIs were calculated following Tierney’s method. Forrest adopted. Potenti adopted.P values were two-sided and a P value less than 0.05 was considered to be statistically significant.To avoid the influence of heterogeneity among these studies, we also conducted a subgroup analysis stratified by different recurrence categories (LR and DR). We defined LR to include true recurrence, local failure, local recurrence or relapse, ipsilateral breast tumor recurrence, locoregional recurrence, and second ipsilateral BC. DR was defined as distant metastasis or failure, and recurrence was defined as any of LR and/or DR, the merger of LR and DR in the same study or not definitely mentioned. All Fifteen studies involving 21,645 participants finally met the inclusion criteria-44. The A total of 20,890 patients (88.3%) received postoperative radiotherapy (RT), including accelerated partial breast irradiation, whole breast irradiation, or MammoSite brachytherapy. Adjuvant systemic therapy (AST) after BCS are defined as hormone/endocrine therapy (HT) only, chemotherapy (CT) only, or both. The directly reported rates range from 0.0% to 77.0% for HT and 0.0% to 66.4% for CT. Excluding studies that did not mention the respective rates of HT or CT, 6,288 patients (49.4%) underwent HT and 4,817 patients (22.5%) accepted CT  are included in analyses.Six articles involved Her-2 typing,34,40-42Triple-molecular subtypes are defined as the combination of dichotomous status of three receptors , including Luminal A (ER + and/or PR + and Her-2-), Luminal B (ER + and/or PR + and Her-2+), Her-2 (ER- and PR- and Her-2+), and triple-negative (ER- and PR- and Her-2-).By setting Luminal A as a baseline, the recurrence risk of patients with Luminal B, Her-2, or triple-negative subtypes are compared by overall and subtotal merge of five studies,35,36,38vs. LA . However in the other seven molecular typing comparisons, the shapes of funnel plots seem symmetrical and the results of Egger’s test do not suggest any publication bias   in non-Hispanic blacks and Asian/Pacific Islanders[BC has become a major cause of morbidity and mortality in women; there aslanders. In addislanders.Currently, local excision combined with adjuvant RT has been well-established as an optimal treatment strategy for early-stage BC,11,50,51Towards this goal, consideration must be given to the combination of underlying pathological and clinical characteristics. Recent studies have shown that molecular subtypes are prognostic for LR and DR after BCS, and immunohistochemical staining is often used to approximately identify these subtypes,19,53. IOur meta-analysis shows that there are significant differences in recurrence risk among various BC subtypes after BCT. First, we analyzed the efficacy of single-molecular subtypes. Although a slightly increased risk has been found in both ER negative and PR negative patients, there is no significant effect on LR after BCT. Only Her-2 positive patients display a significantly higher risk for local and overall recurrence when compared to Her-2 negative individuals Table . It reveRecently, more studies have focused on the combined efficacy of ER, PR, and Her-2 receptors since the 12th St Gallen International Breast Cancer Conference (2011) Expert Panel adopted a new approach to BC classification for therapeutic purposes based on the recognition of intrinsic biological subtypes. TherefoMoreover, the comparison between triple-negative and non-triple-negative subtypes shows the biggest risk difference for overall recurrence (HR = 3.19) and LR (HR = 3.31) among all molecular typing comparisons. When setting Luminal A as a baseline, the HR value of triple-negative subtype (HR = 2.90) remains larger than that of Luminal B (HR = 2.23) or Her-2 subtypes (HR = 2.26) for overall recurrence Table . PreviouSome limitations of this meta-analysis should be acknowledged. First, there are only three studies focused on Asians,36,42 anOur meta-analysis represents a quantified synthesis of all published studies and shows significant differences in recurrence risk among various molecular subtypes after BCT. Her-2 positive and p53 positive subtypes can be considered independent prognostic biomarkers for indicating high LR risk, but triple-molecular biomarkers exhibit higher clinical value than single-molecular biomarkers. Moreover, triple-negative subtype shows the biggest risk for overall recurrence and LR among all molecular subtypes and adjuvant CT should be considered in BCT. Considering the insufficient number of original studies, further research with different ethnicities is needed on this topic, especially the intensity of association between molecular typing and DR risk after BCT.2: I-squared statistic; LR: Local recurrence; PR: Progesterone receptor; RT: Adjuvant radiotherapy.AST: Adjuvant systemic therapy; BC: Breast cancer; BCS: Breast-conserving surgery; BCT: Breast-conserving therapy; CIs: Confidence intervals; CT: Chemotherapy; DR: Distant recurrence; ER: Estrogen receptor; HER-2: Human epidermal growth factor receptor 2; HR: Hazard ratio; HT: Hormone/endocrine therapy; IThe authors declare that they have no conflict of interest.JC, PJ, and SW conceived and designed the study. PJ, J-yZ, and YX collected the data. H-jW contributed to quality assessment. M-hG, BZ, and C-yT performed statistical analyses. JC and H-yC drafted and revised the manuscript. All authors read and approved the final version.


C677T and A1298C are the most common allelic variants of Methylenetetrahydrofolate Reductase (MTHFR) gene. The association between MTHFR polymorphisms and the occurrence of non-alcoholic fatty liver disease (NAFLD) remains controversial. This study was thus performed to examine whether MTHFR mutations are associated with the susceptibility to NAFLD.st, 2016), were searched for eligible case-control studies. The odd radio (OR), 95% confidence interval (CI) and P value were calculated through Mantel-Haenszel statistics under random- or fixed-effect model.A first meta-analysis on the association between the MTHFR polymorphisms and NAFLD risks was carried out via Review Manager 5.0 and Stata/SE 12.0 software. The on-line databases, such as PubMed, EMBASE, CENTRAL, WOS, Scopus and EBSCOhost (updated to April 1P = 0.04) and recessive model , but not the other genetics models, compared with control group. For A1298C, significantly increased NAFLD risks were detected in allele model , homozygote model , dominant model  and recessive model , but not heterozygote model.Eight articles (785 cases and 1188 controls) contributed data to the current meta-analysis. For C677T, increased NAFLD risks were observed in case group under homozygote model (T/T vs C/C, OR = 1.49, 95% CI = 1.03~2.15, T/T genotype of MTHFR C677T polymorphism and C/C genotype of MTHFR A1298C are more likely to be associated with the susceptibility to NAFLD. Human Methylenetetrahydrofolate Reductase (MTHFR) gene is located at chromosome 1p36.3 and contains 11 exons . As a kiNon-alcoholic fatty liver disease (NAFLD), the most common chronic liver disease, is the hepatic manifestation of the metabolic syndrome without a history of excess alcohol consumption –14. The To our knowledge, no meta-analysis on the association of MTHFR genetic variants and overall NAFLD risks has been reported. Hence, we first carried out a meta-analysis to investigate the relationship between MTHFR polymorphisms (C677T and A1298C) and susceptibility to NAFLD. Our finding showed that both C677T and A1298C polymorphisms of MTHFR gene might positively correlate to the risks of NAFLD.st, 2016. There was no language or region restriction. The combinations of following keywords were used: “Methylenetetrahydrofolate Reductase (NADPH)” or “Methylene-THF Reductase (NADPH)” or “MTHFR” or “Methylenetetra hydrofolate Reductase”; “Non-alcoholic Fatty Liver Disease” or “NAFLD” or “Fatty Liver, Nonalcoholic” or “Livers, Nonalcoholic Fatty” or “Nonalcoholic Fatty Livers” or “Nonalcoholic Steatohepatitis” or “Nonalcoholic Steatohepatitides” or “Steatohepatitides, Nonalcoholic” or “Steatohepatitis, Nonalcoholic”; “Polymorphism, Genetic” or “Genetic Polymorphisms” or “Genetic Polymorphism” or “Polymorphism (Genetics)” or “Polymorphisms, Genetic”. The full details of databases searching terms were also provided , Cochrane Central Register of Controlled Trials (CENTRAL), Web of Science (WOS), China National Knowledge Infrastructure (CNKI)/WANFANG, Scopus and EBSCOhost in April 1provided .The eligible case-control studies were identified according to the following inclusion and exclusion criteria. Inclusion criteria: 1) The data on the association between MTHFR polymorphisms and susceptibility to NAFLD was provided; 2) The individual genotype frequencies for MTHFR polymorphisms could be extracted. Exclusion criteria: 1) duplicated studies; 2) reviews or books; 3) non-clinical data; 4) other genes; 5) non-NAFLD diseases; 6) case, trial, or non-polymorphism; 7) meeting/conference abstracts; 8) unavailable data.2 and P value of Hardy-Weinberg Equilibrium (HWE) test in control group. HWE value was calculated by chi-squared test and P value less than 0.05 was considered a departure from HWE.Data was extracted from qualified articles independently by the authors (MYS LZ SLS JNL) using the same reporting form. The controversial evaluations were resolved through discussion. If the data was unavailable, an attempt was made to contact corresponding author to request missing data via E-mail. The following information was extracted: mutation site, first author, year of publication, country, ethnicity, sample sizes in case and control group, source of control, genotyping method, gender and age in case group, disease group, allele and genotype frequencies in each group, The ΧP value, odd radio (OR) and corresponding 95% confidence interval (CI) were calculated by Mantel-Haenszel statistics under the allele, homozygote, heterozygote, dominant or recessive models. P value <0.05 was considered statistically significant association between C677T and A1298C polymorphisms of MTHFR and NAFLD risks. Χ2-based Q statistic and I2 test were applied to analyze the overall heterogeneities. When I2 values < 25% or P value of heterogeneity >0.10, a fixed-effect model was selected for Mantel-Haenszel statistics. Otherwise, a random-effect model was used [The was used –35. Whenwas used . Statistst, 2016), to obtain the eligible case-control studies. Flow chart of studies selection in meta-analysis was shown in We searched the on-line electronic databases, including PubMed, EMBASE, CENTRAL, WOS, CNKI/WANFANG, Scopus and EBSCOhost , EMBASE (n = 29), CENTRAL (n = 0), WOS (n = 24), CNKI/WANFANG (n = 3), Scopus (n = 144) and EBSCOhost (n = 11). After 50 duplicated articles were removed, the 153 articles were excluded by screening the title and abstract: 59 articles are reviews or books; 11 articles do not provide the clinical data; 16 articles are related to the other genes; 46 articles focus on non-NAFLD diseases; 21 articles are case, trial or fail to contain the data of gene polymorphism. 18 potentially articles were then assessed for eligibility. The data was extracted from all these full-text articles. As shown in 2 = 56% and P = 0.004) revealed that high heterogeneity among studies was detected for C677T polymorphism. Random-effect model was thus applied for meta-analysis. The data on the association between C677T allele frequency of MTHFR and susceptibility to NAFLD was obtained . In addition, the potential publication bias was evaluated by Begg’s funnel plot with pseudo 95% confidence limits. The result of P = 0.893) excludes the presence of large publication bias.Next, the genetic association between MTHFR C677T polymorphism and susceptibility to NAFLD was measured. As shown in 2 = 47% and P = 0.02), C/T+T/T vs C/C (I2 = 45% and P = 0.03) and T/T vs C/C+C/T (I2 = 26% and P = 0.16) data indicated the existence of the moderate degree of heterogeneity across studies (2 = 21% and P = 0.21). Pooled analysis for the association between C677T genotype frequencies and the risks of NAFLD was shown in P = 0.04) and recessive model , but not the other genetics models . In addition, the results of HWE test . The subgroup analyses under all genetic models were also performed based on ethnicity or HWE  and HWE P>0.05 subgroup . In order to evaluate the influence of each study on the overall OR under all genetic models, the sensitivity meta-analyses, in which one study is omitted at a time, were also performed. As shown in The contrast of the homozygote model (T/T vs C/C), heterozygote model (C/T vs C/C), dominant model (C/T+T/T vs C/C) and recessive model (T/T vs C/C+C/T) was then detected respectively, through the meta-analysis, in that the data on genotype frequencies of MTHFR C677T polymorphism was available. Genotype distribution and characteristics of MTHFR C677T polymorphism in different case-control studies were shown in  studies . A randoHWE test  in contry or HWE  via StatP>0.05). In addition, all the case-control studies were performed in Caucasian population. We then first performed the meta-analysis between the allele frequency of MTHFR A1298C and the susceptibility to NAFLD under C vs A model. As shown in 2 = 66% and P = 0.001) for meta-analysis. The data  was obtained in C vs A comparison of MTHFR A1298C. The basically symmetric plot  did not provide the statistical evidence for publication bias , A/C vs A/A model (I2 = 53% and P = 0.02), A/C+C/C vs A/A model (I2 = 63% and P = 0.003) was obtained and random-effect model was used. For the C/C vs A/A+A/C model, fixed-effect model was used (I2 = 14% and P = 0.31). A significantly increased NAFLD risks was observed in homozygote model , dominant model  and recessive models , but not heterozygote model . Moreover, similar results were obtained in the sensitivity meta-analyses under all genetic models  and susceptibility to many clinical diseases, such as gastric cancer, hepatocellular carcinoma, NAFLD, neural tube defects, acute lymphoblastic leukemia and renal/heart failure , 38–42. The positive correlation between NAFLD susceptibility and two MTHFR variants (C677T and A1298C) was observed in our statistical evidence. For C677T polymorphism, an increased NAFLD risk was observed under homozygote model (T/T vs C/C) and recessive model (T/T vs C/C+C/T), but not T vs C, C/T vs C/C and C/T+T/T vs C/C models, suggesting that T/T genotype of MTHFR C677T polymorphism might have the increased risks of NAFLD in general population. Moreover, we found that a significantly increased NAFLD risk was detected in Asian population under the comparison of T vs C, T/T vs C/C, C/T vs C/C; C/T+T/T vs C/C. Similarly, the meta-analysis of A1298C polymorphism based on 11 case-control studies in Caucasian population provided the evidence that a significantly increased NAFLD risk was observed under allele model (C vs A), homozygote model (C/C vs A/A), dominant model (A/C+ C/C vs A/C) and recessive model (C/C vs A/A+A/C), but not heterozygote model (A/C vs A/A), suggesting that C/C genotype of MTHFR A1298C polymorphism might be linked to the susceptibility to NAFLD in Caucasian population.The C677T polymorphism means the substitution of C (cytosine) to T (thymine) at nucleotide position 677, which results in the transition from alanine to valine, while A1298C polymorphism refers to the transition of A (adenine) to C (cytosine) at position 1298, which leads to an amino acid substitution from glutamic acid to alanine . Folate There are some shortages or limitations in this meta-analysis, which should be pointed out. For example, no large sample size was included in the case/control groups of meta-analysis. It is still possible that other unpublished or undetected studies are present, although we selected the eligible studies independently according to the inclusion and exclusion criteria. The potential selection bias still may affect the reliability of our findings. Different degree of heterogeneity and departure from HWE was also detected in some comparisons or case-control studies. Furthermore, it was reported that C677C/C1298C compound genotype confers increased risks of NASH in Turkish women patients . UnfortuVery complicated natural history of NAFLD was existed, and multiple genetic or environmental factors contribute to the occurrence and progression of the NAFLD –53. NAFLAll in all, this is the first meta-analysis to provide evidence that C677T and A1298C mutations of MTHFR are significantly associated with an increased risk of NAFLD. The homozygous T/T genotype of MTHFR C677T and C/C genotype of MTHFR A1298C polymorphism seem to be more susceptible to NAFLD. More case-control studies are warranted to validate the conclusion.S1 Table(DOCX)Click here for additional data file.S2 Table(DOCX)Click here for additional data file.S1 Text(DOCX)Click here for additional data file.S2 Text(DOCX)Click here for additional data file.


Low serum Vitamin D is considered to be associated with tuberculosis while the “dangerous” level was not clear. The aim of this study was to identify the association between tuberculosis and serum Vitamin D levels via synthesis of available evidence.A search of EMBASE, Medline, ISI Web of knowledge, and Pubmed was conducted. The number of subjects of tuberculosis and no-tuberculosis groups in four Vitamin D range. Meta-analyses were performed and presented by odds ratios (ORs) and corresponding 95% confidence intervals (CIs).A total of 15 studies involving 1440 cases and 2558 controls were included. A significantly increased risk of tuberculosis was found in two ranges: ≤ 12.5 nmol/L: pooled OR = 4.556, 95% CI = 2.200-9.435; 13-25 nmol/L: pooled OR = 3.797, 95% CI = 1.935-7.405. No statistically significant risk of tuberculosis was found in the range of 26–50 nmol/L . In range 51–75 nmol/L, no positive association was found .This study found that a serum Vitamin D level ≤ 25 nmol/L was significantly associated with an increased risk of tuberculosis while the range of 51–75 nmol/L was not. The range 26-50nmol/L posed potential high tuberculosis risk. Future large-scale, well-designed studies are needed to verify these results. Tuberculosis (TB) remains a major challenge to global public health.  In additVitamin D, an immunomodulatory effector, has been proven to play a critical role in inducing antimycobacterial activity which is accomplished by inhibiting the growth of Mycobacterium tuberculosis (MTB) and up-regulating protective innate host responses., 4 It haA literature search of EMBASE, Medline, ISI Web of knowledge, and Pubmed up to February, 2015 was conducted. Search terms included tuberculosis, TB, serum Vitamin D, Vitamin D, Vitamin D deficiency and cholecalciferol. After obtaining all searching results, we eliminated unrelated publications by title reading and performed manual searches of citations from these original studies. To avoid replicated data in different publications from the same author or the same researching team, we further checked these remaining articles and none of the included studies were found to contain this problem.Studies were included if they meet the following criteria: a) studies must be reported in English and have available full text; b) the organism studied must be Mycobacterium tuberculosis (MTB); c) studies must involve a TB group and a control group, investigate the association between serum Vitamin D concentrations and TB in humans and supply the number of participants in our defined concentration range or quite closing range. Exclusion criteria were: a) reviews or meta-analyses; b) fundamental studies of cells or animal models; c) the studies enrolled patients known to be immunosuppressed . No special limitation of age, population or study design was defined in this study.The Newcastle–Ottawa Scale , a validTwo reviewers (Zeng and Wu) independently collected data from each study. Disagreements were resolved by discussion between them and a third author was consulted to obtain a final decision if a consensus could not be reached between them. The following items were extracted from each study: First author, age of both groups, male percentage of both groups, nationality, ethnicity, source of controls, TB diagnosis standards and methods, number of TB patients and controls in different ranges of serum Vitamin D, study design, percentage of participants in each range, body mass index (BMI) in both groups and the treatment status of participants. Some studies offered detailed figures with obtainable data and we retrieved all available information from these figures. All included studies defined serum Vitamin D by 25-(OH)D3, which has been widely accepted as a better indicator of vitamin D status than 1,25(OH)2D3.  Thus, we2 test and I2 statistic. [All statistical analyses were performed using STATA11.0 software . The strength of association was evaluated by odds ratios (ORs) with corresponding 95% confidence intervals (CIs). All meta-analyses were conducted by using a random-effects model. Heterogeneity among studies was assessed by the χatistic. Subgroup The study selection process is schematically presented in Because these included studies were observational and nonrandomized, the quality of the primary studies were evaluated by the Newcastle–Ottawa Scale.  Five stu2 = 11.9%, P<0.001) and 13–25 nmol/L . Although the pooled OR of levels 26-50nmol/L reached 1.561, the association was not statistically significant. . In the 51–75 nmol/L range, no positive association was found . These results are presented in We obtained the number of subjects in the TB and control groups at different concentration ranges to explore the risk of active TB based on Vitamin D levels. Four concentration ranges were established grounded on original articles: ≤ 12.5 nmol/L, 13–25 nmol/L, 26–50 nmol/L, and 51–75 nmol/L. A significantly increased risk of active TB was identified in serum Vitamin D levels ≤ 12.5 nmol/L . Meta-regression of publication year and mean age of TB and control group also indicated that none of these factors was a substantial source of heterogeneity. Neither Begg’s test nor Egger’s test found publication bias in any analysis.We next conducted the subgroup analyses by design, population, serum Vitamin D analysis method, and source of controls ; however, no significant changes in pooled results or heterogeneity were found. Although pooled OR of level 26-50nmol/L was 1.561, the 95%CI cross 1. Two articles included some participants received anti-TB treatment before vitamin D detection in the concentration range. Considerated the possible influence of anti-TB treatment, we conducted subgroup analysis by treatment status. However, in both untreated group and received treatment group, the results showed no statistically significance in the concentration range.. Further investigation is needed because of the high degree of heterogeneity and relatively weak robustness.A total of 15 studies were included in this analysis. This meta-analysis found that a serum Vitamin D concentration ≤ 25 nmol/L was associated with a significantly increased risk of active TB while the range 51–75 was not associated with active TB. However, the range 26–50 nmol/L was not statistically associated with active TB. Our meta-analysis confirmed the results of a previous meta-analysis that showed that low serum Vitamin D was a risk for active TB and further verified the precise range of low serum Vitamin D posing high risk of TB. To the best of our knowledge, this is the first meta-analysis providing this range.One recent systematic review reported that mean population-level 25(OH)D3 values varied considerably across the studies (range: 4.9 to 136.2 nmol/L), and more than one-third of the studies reported mean values < 50 nmol/L. In regarOur results demonstrated lower serum Vitamin D level was associated with higher the risk of active TB. Only serum Vitamin D between 12.5 and 25 nmol/L (95% CI: 1.935–7.405) and ≤ 12.5 nmol/L (95% CI: 2.200–9.435) were associated with a significantly increased risk of active TB. However, 51–75 nmol/L (95% CI: 0.708–1.900) was identified as not associated with an increased risk of active TB. Influence analysis and publication bias analysis confirmed that these results were reliable and robust.P = 0.051). Some original articles included in our study suggested <50nmol/L was the risk level [It is interesting that serum vitamin D between 26 nmol/L and 50 nmol/L was not statistically associated with an increased risk of TB according to our analysis. Although the pooled OR was 1.561 in the range 26–50 nmol/L, the 95% CI was cross 1.0 Click here for additional data file.S1 AppendixThe scoring criteria of the original studies included in our meta-analyses.(DOC)Click here for additional data file.


Insulin-like growth factor-1 (IGF-1) is reported to be neuroprotective in the setting of Parkinson’s disease (PD), and there is increasing interest in the possible association of serum IGF-1 levels with PD patients, but with conflicting results. Therefore, we conducted a meta-analysis to evaluate the association of serum IGF-1 levels in de novo, drug naïve PD patients compared with healthy controls.Pubmed, ISI Web of Science, OVID, EMBASE, and Cochrane library databases from 1966 to October 2014 were utilized to identify candidate studies using Medical Subjective Headings without language restriction. A random-effects model was chosen, with subgroup analysis and sensitivity analysis conducted to reveal underlying heterogeneity among the included studies.In this meta-analysis, we found that PD patients had higher serum IGF-1 levels compared with healthy controls . Subgroup analysis demonstrated that the source of heterogeneity was population differences within the total group. Sensitivity analysis showed that the combined MD was consistent at any time omitting any one study.The results of this meta-analysis demonstrate that serum IGF-1 levels were significantly higher in de novo, drug-naïve PD patients compared with healthy controls. Nevertheless, additional endeavors are required to further explore the association between serum IGF-1 levels and diagnosis, prognosis and early therapy for PD. The morbidity of this chronic progressive disorder is anticipated to rise as the affected population continues live longer and increase in number. Althoughell death and augmell death which maell death Indeed, ell death–13 ThereOur study was performed according to the Preferred Reporting Items for Systematic Reviews and Meta-analyses (PRISMA) S1 Chec. We searThe eligibility of articles included in this meta-analysis were evaluated by the following inclusion criteria: (1) case-control studies comparing serum IGF-1 levels between de novo, drug-naïve idiopathic PD patients and healthy controls, or cohort studies with detailed baseline data; (2) the diagnosis of PD must be made according to the UK Parkinson’s Disease Society Brain Bank; (3) detReview articles, commentaries, and conference proceedings without new data were excluded from this meta-analysis. Additionally, all articles pertaining to the fundamental pathophysiology of IGF-1 as related to the pathogenesis or treatment of PD were excluded. Other studies were excluded if their participants were diagnosed with secondary parkinsonism, atypical parkinsonism, familial parkinsonism or suffered from any disease that may affect serum IGF-1 levels. Two authors independently screened all articles based on the above exclusion criteria.The Newcastle-Ottawa Scale (NOS) criteria, includinInitially, 35 studies were retrieved with MeSH words: “Parkinson’s disease” and “insulin-like growth factor-1”. After reviewing the titles and abstracts, 9 replicate studies, 8 fundamental research papers, 2 reviews and 6 other articles irrelevant to serum IGF-1 levels in PD patients were eliminated. Of the remaining 10 papers, 2 commentary papers, 1 article with no de novo drug-naïve PD patients, 1 article with plasma IGF-1 and 1 study investigating the IGF-1 gene were excluded. Ultimately, 5 studies with 166 PD patients and 323 healthy controls were included in our meta-analysis. Of these 5 studies, 2 were conducted in Germany, 2 were carried out in Italy and 1 was conducted in Japan. The selection process is further detailed in The Newcastle-Ottawa Scale (NOS) criteria is composed of 3 major categories: selection (4 items), comparability (1 item) and exposure (3 items). Each item in the selection and exposure section can be awarded a maximum of one point, with a maximum of two points in the comparability section. A higher score indicates higher quality methodology.The random-effects model was used to calculate the pooled effect size of the five studies. All available data from the five included studies were used to compare different serum IGF-1 levels between de novo, drug naïve PD patients and healthy controls. Notably, a significant difference was found between the two groups, and the pooled mean difference in the random-effects model was 17.75, 95%CI: 6.01, 29.48 .2 = 81%, P<0.0001). Therefore, we divided the total group into three subgroups according to the nation where the research was performed: Germany (N = 2), Italy (N = 2), and Japan (N = 1). These three subgroups resolved the overall heterogeneity, with study heterogeneity no longer significant in the Germany subgroup  or the Italy subgroup . Heterogeneity analysis is not applicable to the single study in the Japan subgroup. The various subgroups are represented by forest plots in Sensitivity analysis was performed to evaluate the stability and reliability of our meta-analysis by repeating analysis in various combinations of the included studies in which a single study is omitted during each reanalysis. No substantial alteration of the results were found in this sensitivity analysis, with the combined mean difference (MD) ranging from 12.70  via omission of the research by Godau 2010, to 22.72  via omission of the research by Numao 2014, indicating statistically robust results. Next, subgroup analysis was carried out to assess for heterogeneity within this meta-analysis. Due to variations among different populations’ serum IGF-1 levels, this meta-analysis is highly heterogeneous Click here for additional data file.