CAT C262T, GPX1 Pro198Leu, MnSOD Ala16Val, GSTM1, GSTT1, and GSTP1 Ile105Val polymorphisms and the risk of CML. No association was observed between CML and variant genotypes of GPX1, MnSOD, GSTM1, and GSTT1 polymorphisms in any of the investigated cases. Our study suggests that the homozygous variant genotype of the GSTP1 Ile105Val gene polymorphisms may be associated with the risk of developing CML , while the heterozygous genotype of the CAT C262T polymorphism seems to have a protective effect against CML . In most cases, no association was found between laboratory parameters and prognostic factors and the variant genotype of investigated gene polymorphisms. We concluded that CAT, GPX, MnSOD, GSTM1, and GSTT1 gene polymorphisms are not associated with the risk of CML. Variant genotype of the GSTP1 Ile105Val gene polymorphisms may contribute to the risk of developing CML.Oxidative damage at the DNA level may be promoted by high levels of reactive oxygen species (ROS), leading to genomic instability and increased neoplastic risk. Superoxide dismutase (SOD), glutathione peroxidase (GPX), and catalase (CAT) enzymes are implicated in the prevention of DNA damage by ROS. The aim of the study was to investigate the relationships between  According to Yuzhalin and Kutikhin  high lev2O2. Hydrogen peroxide is decomposed by mitochondrial peroxiredoxins . However, some of the polymorphisms of the MnSOD gene lead to an increased superoxide dismutase activity, yielding increased quantities of H2O2. This process could contribute to the neoplastic transformation of the cells [ MnSOD gene , located in the mitochondrial targeting sequence, has been described to be associated with the overexpression of MnSOD, leading to an increased production of hydrogen peroxide (H2O2), which generates increased levels of ROS, if not subsequently neutralized [Manganese superoxide dismutase (MnSOD) catalyzes the conversion of highly reactive superoxide radicals to Hhe cells . Such a 2O2 to water and molecular oxygen [ CAT gene, namely C262T, has been associated with lower catalase activity and thus increased levels of ROS [Catalase (CAT) participates in defense mechanisms against oxidative stress by converting Hr oxygen . A commo2O2 and other organic peroxidases to water with reduced glutathione [ GPX1 gene, namely Pro198Leu, has been shown to be associated with malignancies [ GPX1 gene polymorphism may increase the risk of cancer induced oxidative DNA damage.The antioxidant enzymes glutathione peroxidases (GPX) are implicated in protection of cells against oxidative damage by reducing Htathione . A polymgnancies . It was gnancies  in a cel GST genes, are implicated in detoxification of xenobiotics, and they have a crucial role in protecting tissue from oxidative damage. The variation at the GSTM1, GSTT1, and GSTP1 loci has been linked with an increase in the occurrence of cancers, probably due to an increased susceptibility to environmental toxins and carcinogens [ GST genes: the homozygous deletions of the GSTM1 and GSTT1 genes (the “null genotypes”) and the Ile105Val substitution in the GSTP1 gene. It has been assumed that these polymorphisms lead to reduced enzymatic activity, thus decreasing their detoxification ability for certain mutagens and carcinogens. This could cause in turn DNA damage and thus a greater risk of developing cancers [Glutathione S-transferases (GSTs) enzymes, encoded bycinogens . The res cancers . CAT C262T, GPX1 Pro198Leu, and MnSOD Ala16Val gene polymorphisms and the risk of CML has not been previously investigated.To our knowledge, the relationship between the GSTT1, GSTM1, and GSTP1 Ile105Val polymorphisms and the susceptibility to chronic myeloid leukemia (CML) was investigated by different studies, but with conflicting results [The association between results –12. CAT C262T, GPX1 Pro198Leu, MnSOD Ala16Val, GSTM1, GSTT1, and GSTP1 Ile105Val polymorphisms and the risk of CML.The aim of the present study was to investigate the relationships betweenThe research protocol was approved by the Ethics Committee of the University of Medicine and Pharmacy Tirgu Mures, Romania, and written informed consent was obtained from controls and CML patients. The study was conducted in accordance with the Declaration of Helsinki.Our study consisted of 489 individuals from the same geographical area , which were divided into two groups: the control and the patients groups.The control group included 321 healthy individuals without a history of any type of cancer . BCR-ABL fusion gene by reverse transcription polymerase chain reaction.The patients group was composed of 168 previously untreated adult CML cases  diagnosed at the Hematology Clinic from Tirgu Mures and Hematology Department of the “Ion Chiricuta” Cancer Institute from Cluj-Napoca, Romania, between December 2010 and April 2014. CML was diagnosed based on detection of Philadelphia chromosome by standard cytogenetic and/orAccording to the clinical phases of the disease we included our CML patients into two groups as follows: 141 cases in the group with patients in chronic phase and 27 (16.1%) cases in the group with patients in advanced phase (accelerated and blast crisis).Additional chromosomal abnormalities (ACA) were found in 22 (13.1%) of our CML cases.The frequency of ACA in CML patients in advanced phase (accelerated and blast crisis) of the disease was 59% (13 cases).Peripheral blood samples were collected in sterile tubes containing EDTA and genomic DNA was isolated using the Quick-gDNA MiniPrep kit  and Wizard Genomic DNA Purification Kit  according to the manufacturer's instructions.CAT C262T, GPX1 Pro198Leu, MnSOD Ala16Val, and GSTP1 Ile105Val genotyping was performed using polymerase chain reaction and restriction fragment length polymorphism (PCR-RFLP) assays as previously described [escribed , 13–15.GSTM1 and GSTT1 polymorphisms were analysed using a multiplex polymerase chain reaction as previously described [escribed . t-test, Mann-Whitney test, ANOVA test, or Kruskal-Wallis test, when appropriate. We used the Bonferroni correction or Dunn's correction in order to account for multiple comparisons. The level of statistical significance was set at P value < 0.05.Statistical analysis was performed using the Statistical Package for Social Sciences . Data were considered as nominal or quantitative variables. Nominal variables were characterized using frequencies. Quantitative variables were tested for normality of distribution using Kolmogorov-Smirnov test and were characterized by median and percentiles (25–75%) or by mean and standard deviation (SD), when appropriate. A chi-square test was used in order to compare the frequencies of nominal variables. Quantitative variables were compared using  CAT 262T allele in CML patients was 22.9%, while in controls it was 27.1% . The allele frequency of the variant Leu allele for GPX1 Pro198Leu polymorphism was 50.4% in CML group and 51.5% in control group . The frequency of the MnSOD 16Val allele was 54.7% in the patient group and 57.8% in the control group . The allele frequency of GSTP1 105Val was 22.9% in CML patients and 17.4% in controls . The frequency of GSTM1 and GSTT1 null genotypes was 59.5% and 20.8%, respectively, in CML group, and 59.2% and 20.9%, respectively, in the control group . In addition we investigated the combined effects of GSTM1 and GSTT1 null genotypes regarding the CML risk. There were 18 patients (10.7%) and 45 (14%) controls harbouring both GSTM1 and GSTT1 null genotypes, which means a similar distribution .The frequency of the variant GSTP1 Ile105Val polymorphism was seen significantly more frequently in patients than in controls .When analysing the genotypes of the six polymorphisms genotyped, we concluded that the variant homozygous of the CAT C262T polymorphism in heterozygous state was found to be protective against CML .Interestingly, theP value > 0.05 for all these comparisons). The genotype distribution of the six  gene polymorphisms in patients and controls is presented in detail in All the genotypes for all other polymorphisms analysed shared similar frequencies between patients and controls , Hasford and Sokal risk groups, the EUTOS score, the WBC count, blasts, and basophils. No association was observed between the laboratory parameters  and the CAT C262T, GPX1 Pro198Leu, MnSOD Ala16Val, GSTM1, GSTT1, and GSTP1 polymorphisms investigated had no impact on the risk of CML with respect to gender  and age .The GPX1 Pro198Leu, MnSOD Ala16Val, GSTM1, GSTT1, and GSTP1 was associated with Hasford and Sokal risk groups and EUTOS score in CML patients. In the case of CAT C262T polymorphism, it was not associated with Hasford risk groups and EUTOS score, but it was associated with Sokal risk group  in CML patients.None of the polymorphismsGPX1 Pro198Leu, MnSOD Ala16Val, CAT C262T, GSTM1 and GSTT1 null genotypes, and GSTP1 Ile105Val) in patients stratified after the clinical phases of the disease (one group was represented by chronic phase and the other group was represented by accelerated or blast phase). None of the polymorphisms analyzed was associated with the clinical phases of CML .We analysed the distribution of the six polymorphisms .Also, we performed a comparison between patients with additional chromosomal abnormalities (ACA) and variant genotypes for the investigated polymorphisms. The frequency ofGSTT1 null genotype, GSTP1 Ile105Ala, CAT C262T, GPX1 Pro198Leu, and MnSOD Ala16Val) and the occurrence of ACA in CML patients .In the current study, no association was observed between the other five polymorphisms analyzed .In the present study we observed no significant association between the GPX1 Pro198Leu polymorphism is associated with an increased risk of developing bladder [ GPX1 Pro198Leu polymorphism and prostate cancer risk [It has been reported that the bladder , 17, bre bladder , and lun MnSOD, CAT, and GPX1 are involved in the same biological pathways, we analysed a possible synergetic effect of the polymorphisms involving the genes coding for these enzymes. The combined variant (heterozygous and homozygous) genotypes for the CAT C262T, GPX1 Pro198Leu, and MnSOD Ala16Val gene polymorphisms revealed no association with the risk of CML . The frequency of the combined variant genotypes of the CAT C262T, GPX1 Pro198Leu, and MnSOD Ala16Val polymorphisms was 33.3% (56 cases) in CML group and 40.05%  in control group.Taking into account that GSTM1 and GSTT1 null genotypes were not associated with the risk of CML. These findings are in agreement with the study conducted by Löffler et al. [ BCR-ABL positive CML cases and 150 random, healthy blood donors from Germany.In the present study, we observed that ther et al.  on 141 B GSTT1 null genotype and the risk of CML was reported by Özten et al. [ GSTM1 and GSTT1 null genotypes were significantly associated with the risk of CML [A positive association betweenn et al. , Taspinan et al. , and He n et al. . The comk of CML , 11. Simk of CML .A recent meta-analysis performed by He et al.  suggesteWe did not observe any association between combined GSTM1 and GSTT1 null genotypes and the risk of CML. Our findings are different compared to those reported by Bhat et al. and He et al. , 9. The  GSTT1 null genotype was associated with increased risk of AML in a Chinese population and no association between GSTM1 null genotype and GSTP1 Val/Val genotype and AML as independent factors [Zhou et al. reported the results of a case-control study conducted on 163 AML cases and 204 controls and found that factors . GSTM1, GSTT1, and GSTP1 polymorphisms have a reduced enzyme function. Therefore, they may potentially have an increased cancer susceptibility due to a decreased ability to detoxify carcinogens and reactive oxygen species [ GSTT1 null, GSTM1 null, and GSTP1 Val/Val genotypes and CML risk. In addition, we investigated the effects of combined GSTM1 or GSTT1 null genotype with variant GSTP1 genotypes on CML risk. We failed to find an association between variables mentioned above.According to Zhou et al., individuals bearing species . Our fin GSTP1 Ile105Val is a risk factor for CML . Similar results were described in 260 CML cases from Hyderabad, India, by Sailaja et al. [ GSTP1 Ile105Val genotypes and CML patients in accelerated and blast crisis.The current study indicated that the presence of homozygous variant genotype  GSTT1, GSTM1, and GSTP1 Iie105Val polymorphisms are presented in The findings of different studies regarding leukemic risk and CAT C262T, GPX1 Pro198Leu, MnSOD Ala16Val, GSTM1, GSTT1, and GSTP1 polymorphisms in our patients with CML and controls. Our results suggest no effects of the variant combined genotypes of all six investigated gene polymorphisms and the risk of CML. Similar results were obtained in CML cases with three, four, and five variant genotypes from all the combined CAT C262T, GPX1 Pro198Leu, MnSOD Ala16Val, GSTM1, GSTT1, and GSTP1 polymorphisms.We also studied the combined genotype distribution of the investigated GPX1 Pro198Leu, MnSOD Ala16Val, GSTM1, and GSTT1 gene polymorphisms are not associated with the risk of CML. Our study suggests that the homozygous variant genotype of the GSTP1 Ile105Val gene polymorphisms may contribute to the risk of developing CML, while the heterozygous genotype of the CAT C262T polymorphism seems to have a protective effect against CML.In conclusion, our data suggests that


The previous genome-wide studies have shown that rs7193343 single-nucleotide polymorphism (SNP) in zinc finger homeobox 3 (ZFHX3) gene correlate with risk of atrial fibrillation (AF). However, the distribution of this SNP differs significantly among various populations. The present study was to investigate whether combined evidence shows the association between ZFHX3 rs7193343 SNP and the risk of AF in various populations.A systematic search of all studies published through Dec 2014 was conducted using the Medline, Embase, WanFang, ScienceDirect, CNKI, and OVID databases. The case-control studies that evaluated an association between rs7193343 SNP and risk of AF were identified. The association between the ZFHX3 rs7193343 SNP and AF susceptibility was assessed using genetic models.We collected 10 comparisons from six studies for rs7193343 SNP, including 1037 cases and 4310 controls in Asian, 5583 cases and 38215 controls in Caucasian, and then performed an updated meta-analysis and subgroup analysis based on ethnicity. In overall population, the occurrence of AF was found to be associated with T-allelic of rs7193343 SNP in ZFHX3 . In subgroup analysis, we observed there was significant association between T-allele of rs7193343 and risk of AF in Caucasian subgroups , but no statistically significance  in Asian population.In Caucasian population, genetic variant rs7193343 SNP is associated with risk of AF in Caucasian population. However, no association is found in Asian population based on the current evidence. Further studies with larger sample size involving case-control populations with multiple ethnics are still required in the future. Atrial fibrillation (AF) is the most common sustained cardiac arrhythmia in clinical practice, and affect individuals suffer from increased rates of stroke, and lead to higher risk of incidence and mortality of cardiovascular disease . The incRecently, studies have provided unequivocal evidence that AF has the important relevance of genetic factors . AF has Therefore, the aim of the present meta-analysis was to investigate whether combined evidence shows the association between ZFHX3 rs7193343 polymorphism and the risk of AF in various populations, determining whether there was heterogeneity among the studies.We performed a systematic search of Medline, Embase, WanFang, ScienceDirect, CNKI, and OVID to identify published epidemiological studies through Dec 2014 that were related to the rs7193343 ZFHX3 polymorphism and AF. The medical subject headings  “zinc finger homeobox 3”, “genetic polymorphism”, “atrial fibrillation”, and the free-text words “ZFHX3” or “rs7193343” were combined. Only studies published in English or in Chinese were included in the present study. Furthermore, the reference lists of all of the full text papers were examined to identify any initially omitted studies. Secondary searches of the grey literature were not performed.Articles from peer-reviewed medical journals were included if they reported on studies using case-control, nested case-control, cross-sectional design, or cohort and provided sufficient data to calculate an odds ratio (OR) and corresponding 95 % confidence interval (CI). After that, comparisons of laboratory methods and overlapping study data were excluded. Using Hardy-Weinberg equilibrium (HWE), we excluded those studies that contain genotype frequencies that weren’t meet criteria in the control groups.Two reviewers independently screened the titles and abstracts for the eligibility criteria. Subsequently, reviewers read the full text of the studies that potentially met the inclusion criteria, and the literature was reviewed to determine final inclusive data. If inclusions have disagreements, we reached a consensus through discussion.With a standard data extraction form used, two of the authors independently extracted the following data from each full-text report. The data extracted from the candidate studies included the title, authors, published year, number of cases or controls, ethnicity, age, gender, study design, genotyping method, genotype distribution, and frequency of T-allele of the rs7193343 polymorphism in cases or controls. We also examined whether the genotype distributions of the control groups followed HWE .Data analysis was conducted using STATA 12.0 . The association between the ZFHX3 rs7193343 polymorphism and AF susceptibility was assessed under the following genetic models, which were treated as a dichotomous variable, T-allele versus C-allele for allele level comparison.2, was used to quantify the inconsistency among the studies’ results with values of 50 % or higher and the large heterogeneity for values of 75 % or higher [Between-study heterogeneity was tested using Q statistics, and P < 0.1 was considered statistically significant. The Mantel-Haenszel method for fixed effects and the Der-Simonian and Laird method for random effects were used to estimate pooled effects . We usedr higher . The datPublication bias was conducted both visually by using a funnel plot and statistically via Begg funnel plots and Egger’s bias test, which measures the degree of funnel plot asymmetry , 23. TheWe initially obtained 311 potential articles, and after screened abstract, among which most were excluded for no relevance to our analysis. Eleven articles then were removed because small number cases and unusable data. Finally, five studies , 24, 25 A total of 10 comparisons from six studies for rs7193343 polymorphism were involved in presence updated meta-analysis containing 6620 cases and 42525 controls  (Table 2 = 0.0 %). These results indicated that there was statistically significant in the association between rs7193343 and AF.As shown in Fig. 2 = 0.0 %). Three studies reported the association between rs7193343 polymorphism and AF in Asian population, we used a fixed-effects model analysis because there was no heterogeneity among three studies , shown no statistically significance    represents the most serious sustained arrhythmia in the process of clinical diagnosis and treatments. It is the feature of uncoordinated atrial activation. Some studies shown that AF was a important factor with a lifetime risk of one in four for men and women 40 years of age and older . CurrentRecently, a growing body of researches provided the relationship of a variant in the ZFHX3 gene on chromosome 16q22, rs7193343-T and AF in multiple population of different ancestry. However, their conclusions were not consistent. The distribution of genotype of rs7193343 polymorphism in ZFHX3 gene differed significantly among populations. Gudbjartsson et al.  expandedThe present study demonstrated that there was statistically significant in the association between T-allele frequency of rs7193343 and AF in overall population. Interestingly, subgroup analysis was performed according to the ethnicity, showing significant association between T-allele frequency of rs7193343 and AF in Caucasian population but not in Asian population. Although a previous meta-analysis was conducted recently with positive results , no subgRs7193343 is an intronic SNP located in the zinc finger homeobox 3 (ZFHX3) gene (chromosome 16q22) . The varThe following potential factors may be account for the different pooled results between the above two ethnical group: (1) inconsistence in genetic background , (2) difAlthough the results of test for heterogeneity were not significant for subgroup population, heterogeneity cannot be completely resolved. The heterogeneity of included studies includes the following: quality of included studies, gender proportion, age, classification of AF, different genetic background, environmental factors, sampling criteria and cultural difference. The heterogeneity of genetic effects between individual studies may also be caused by the existence of gene-environmental or genetic interaction. In spite of only ten comparisons included, the pooled results may have clinical significance according to the current evidence. Although no publication bias was found in the present study, it is also important to bear in mind that publication bias may exist, since the negative results are less likely to be published. Accordingly, while the results of this meta-analysis should be considered appropriate, the above methodological defects should be considered when interpreting the findings. The present study could provide objective evidence by using genetic meta-analysis approach.The primary limitations of this updated meta-analysis include the following: (1) the statistical efficacy could be improved by including more studies; (2) Some non-English publications not being included in this study may have caused important studies to be overlooked and publication bias from significant conclusions being more easily published; (3) The genotyping data of SNPs rs7193343 were insufficient, so we were unable to conduct meta-analysis based on the limited studies data; (4) the combined analysis of the results may be affected by the impact of clinical heterogeneity, including the general condition of the cases, habits, medical history, medication compliance and other factors.Genetic variant rs7193343 is associated with a crucial risk of AF in Caucasian population, whereas SNP rs7193343 is not association with AF in Asian population. This variant could have a clinical utility in estimating AF risk. On the other hand, more studies with great sample size and multiple ethnics are required to determine the significance of rs7193343 SNP between them and to define the effect of the size of T-allele on susceptibility to AF.Ethical approval was not required for this study design.


Several studies on the association of tumor necrosis factor alpha (TNF-α) polymorphisms with recurrent pregnancy loss (RPL) risk have reported conflicting results. The present meta-analysis was conducted to provide a more precise estimation of these relationships and to investigate the real association between TNF-α polymorphisms and RPL.An extensive eligible literature search for relevant studies was conducted on PubMed, Embase, and The Cochrane Library from their inceptions to May 12, 2015. Specific inclusion criteria were used to evaluate articles. The odds ratio (OR) with 95% confidence intervals (CIs) were used to assess the strength of associations. Statistical analyses were performed by the STATA12.0 software.10 case–control studies including 1430 RPL patients and 1727 healthy controls were identified. Meta-analysis indicated that TNF-α-308G/A (rs1800629) polymorphism in the TNF-α gene correlated with elevated RPL risk whereas no significant association was observed between TNF-α-238G/A (rs361625) and RPL.The current meta-analysis demonstrates that TNF-α-308G/A polymorphism in the TNF-α gene is associated with susceptibility to RPL. Recurrent pregnancy loss (RPL) is defined as three or more consecutive spontaneous abortions before the 20th week of gestation , 2. It iSome studies have led to the awareness that these unexplained RPL might be due to dysregulated immunologic factors , 8. ConsAn increasing number of genetic association research are conducted to determine the genetic background of RPL . ResearcAs stated above, several original studies have reported the correlations between TNF-α polymorphism and RPL, but the results are unconvincing and unreliable, which may partly be due to the relatively small samples and different human populations. A previous meta-analysis was conducted in 2012 trying to investigate this relationship . In viewArticle searches were performed independently by two investigators and the final search strategies were performed with agreement. An extensive systematic literature search for relevant studies was conducted with PubMed, Embase, and The Cochrane Library from their earliest available date through May 12, 2015. For TNF-α polymorphisms and RPL risk, the search terms were as follows: (“tumor necrosis factor” OR “TNF”) AND (“recurrent pregnancy loss” OR “recurrent spontaneous miscarriage” OR “recurrent spontaneous abortion”) AND “polymorphism”. All the articles about three or more miscarriages associated with TNF-α polymorphism were included. Moreover, all articles were published in the primary literature to avoid duplicating analyses. All clearly irrelevant studies, editorials, case reports, and review articles were excluded. Furthermore, literatures cited in the reference sections of review articles and other relevant studies were searched manually for additional eligible studies.Eligible studies were selected according to the following explicit inclusion criteria:(1) the original study was designed as an independent genotyped case–control study; (2) inclusion of both RPL cases and non-RPL controls; (3) investigation of the correlation between TNF-α genetic polymorphisms and RPL risk; (4) adequate data that could be used to calculate the numbers of genotype frequency had to be clearly described in the original study. In addition, the following exclusion criteria were also used: (1) no healthy control population and raw data; (2) genotype frequency unavailable; (3) non-conformity with the criteria for RPL; and (4) duplication of previous publications.The bibliographic search and data extraction were conducted independently by two investigators from all eligible publications according to the above inclusion criteria. Any disagreement was subsequently resolved by consensus with a third author. The following characteristics was collected prospectively: the first author’s name, year of publication, source of publication, country of origin, genotype number in cases and controls, genotype method, and gene polymorphism Table .Table 1MDescription of the case and control subjects’ characteristics ;Assessment and validation of miscarriage in the patients . Adequate validation would include confirmation by scan or pathological examination; inadequate validation would include recollection of the patient as the only evidence or a biochemical pregnancy without ultrasound evidence of pregnancy;Description of the laboratory procedures for the genotyping ;Elimination of confounding factors in patients . “Adequate” refers to the elimination of the proven causes of recurrent miscarriage ;Equal assessment for confounding factors in the case and control groups .The quality of the included studies was assessed according to the following criteria from the previous report .Descript2 statistic. When P value of Q test was less than 0.05 and/or I2 more than 50%, it was considered there was significant heterogeneity and a random-effects model was used, otherwise, the fixed effects model was selected. Potential publication bias was diagnosed statistically via the funnel plots and Egger’s tests. Moreover, subgroup analyses were conducted to explore reasons for heterogeneity. In all analyses, two-sided P value <0.05 was considered to be statistically significant.Data management and analysis were performed using the programs STATA version 12 . Crude odds ratios (ORs) with 95% confidence intervals (CIs) were used to assess the association between the TNF-α polymorphisms and the risk of RPL. Analysis of polymorphisms was conducted in at least three studies. The ORs was calculated for the allele model, homozygote comparison, heterozygote comparison, dominant model, and recessive model based on the genotype frequencies in cases and controls. Heterogeneity was evaluated with Cochran's Q test and IWe identified 19 articles that evaluated the association between TNF-α gene SNPs and RPL risk. 10 of these studies met the eligibility criteria defined in the Materials and Methods section , 24–28. The 10 studies reported here were published between 2001 and 2013 and involved totals of 1430 cases and 1727 controls. Overall, 5 studies were conducted in Asia, 2 in South America, 2 in Europe, and 1 in Africa. Among all the SNPs of the TNF-α gene addressed, −308G/A, and-238G/A were the most common. The DNA samples were extracted from blood in the included studies. Methods used for genotyping include direct DNA sequencing and polymerase chain reaction-restriction fragment length polymorphism (PCR-RELP). The main characteristics of all the included studies are summarized in Table P = 0.002)      Fig. , the het17) Fig.  and the 03) Fig. , but no 57) Fig. , AA + GA6) Table .Fig. 2FoP < 0.001), 0.345 , and 2.759  for Asians under codominant, homozygous and recessive genetic models respectively. The results of this subgroup analysis are displayed , AA vs. GG , AA vs. GA , AA + GA vs. GG , and AA vs. GA + GG   and −308 G/A .Funnel plots and the Egger's test were employed to assess the publication bias of included studies. In the funnel plot analysis, the shape of funnel plot seemed symmetrical Fig. . FurtherRPL is a common disorder and represents a major concern for reproductive problem  affectinAs a pleiotropic cytokine, TNF-α has attracted attention because of its involvement in the promotion of inflammatory response, autoimmune, endocrine and neoplastic diseases. Furthermore, there is convincing evidence that TNF-α induces apoptosis of cytotrophoblasts, which suggested that aberrant expression of TNF-α may have harmful effects on placental development and function . IncreasThe incidence of RPL is controlled by genetic factors, and genetic polymorphisms have been associated with poor pregnancy outcome . In the The present meta-analysis included 1430 cases and 1727 controls from 10 independent case–control studies. The results suggested that -308G/A polymorphisms related with an elevated risk of RPL, indicating that -308G/A may be risk factors for RPL. However, no statistically significant association was observed between -238G/A and RPL risk. One possible reason behind this pattern of results could be that -308G/A polymorphism were more impactful than -238G/A on TNF-α gene expression and protein production, thereby possibly contributing to RPL risk. Moreover, stratification by geographic position, the polymorphism of -308G/A was significantly associated with RPL risk for Asians rather than non-Asians. The reason for the discrepancy is unclear, but it might be explained in part by geographic variation in the frequency of the allele A, i.e., TNF-α-308G/A polymorphism in the Asian patients was higher than non-Asians (14.86 vs.9.63%). The difference could also be explained by the small sample sizes of several included studies, which may lead to substantial errors from estimation . Thus, TIn addition, no correlation was observed between -238G/A polymorphism and susceptibility to RPL. Smaller studies are often characterized by larger effects in this meta-analyses of -238G/A polymorphism, which can be possibly explained by publication bias . It is pOur meta-analysis of the relationship between the TNF-α polymorphisms and RPL risk differs from the results previously reported by Zhang . The preAs with other meta-analyses, it was prudent to acknowledge that several potential limitations were apparent in this analysis. First, the number of studies and the sample sizes were relatively small for analysis of each gene polymorphism thereby having insufficient power to estimate the association between TNF-α genetic polymorphisms and RPL risk. Second, a meta-analysis is a retrospective study, the selection bias would lead to the heterogeneity of the results, and thereby possibly influencing the reliability of our conclusions. Even though the studies have similar inclusion criteria, there are also some differences such as different examinations of each group patients, potential confounders  might skew the results. Finally, our study only included articles published in English from the three selected databases, which might limit the results of the meta-analysis. It was critical that larger and well-designed studies should be performed to reevaluate the association precisely.In summary, this meta-analysis systematically evaluated the association between TNF-α genetic polymorphisms and RPL risk and demonstrates that -308G/A polymorphism in the TNF-α gene is associated with susceptibility to RPL. This polymorphism might be a risk factor for RPL. Further functional studies between TNF-α gene and RPL risk are warranted.


LILRA3 gene in MS have been repeatedly reported with inconsistent results.Multiple sclerosis (MS) is a neurodegenerative, autoimmune disease of the central nervous system. Genome-wide association studies (GWAS) have identified over hundred polymorphisms with modest individual effects in MS susceptibility and they have confirmed the main individual effect of the Major Histocompatibility Complex. Additional risk loci with immunologically relevant genes were found significantly overrepresented. Nonetheless, it is accepted that most of the genetic architecture underlying susceptibility to the disease remains to be defined. Candidate association studies of the leukocyte immunoglobulin-like receptor LILRA3 alleles were discriminated in a single-tube PCR amplification and the resulting products were visualized by their different electrophoretic mobilities.In an attempt to shed some light on these controversial findings, a combined analysis was performed including the previously published datasets and three newly genotyped cohorts. Both wild-type and deleted LILRA3 deletion , even after stratification by gender and the HLA-DRB1*15:01 risk allele.Overall, this meta-analysis involved 3200 MS patients and 3069 matched healthy controls and it did not evidence significant association of the  HLA-DRB1*15:01 allele has been described as the main genetic susceptibility factor for the disease, with modest evidence of heterogeneity (I2 = 25%). Subsequently, a combined analysis of the previously published datasets adding the data of the three newly genotyped independent cohorts was performed. Overall, this meta-analysis including 3200 MS patients and 3069 matched healthy controls did not reveal significant evidence for association and yielded an effect size close to one  with moderate evidence for heterogeneity (I2 = 32%). The number of individuals included would allow the detection of an effect with OR> 1.17, with a statistical power over 80%. A further sensitivity analysis was performed in this study, which provided I2 values ranging from 2% to 41%: the analysis excluding the dataset “2009-Spain (Madrid)” led to the lowest heterogeneity, and again statistical significance was not observed (p = 0.49).Four previously published reports have studied the association of the datasets . Meta-anLILRA3 deletion and the classical MS risk factor HLA-DRB1*15:01, a meta-analysis of those datasets including both variants was performed  = 1.03 (0.88–1.20)].To check for an epistatic interaction between the erformed . The metLILRA3 deletion/ HLA-DRB1*15:01 double-positive subjects and again, a significant difference was not detected . Neither significant evidence of sex-bias was observed in carriers of the LILRA3 deletion lacking HLA-DRB1*15:01 (data not shown).Further analysis was performed stratifying by gender the The role of genetics in MS susceptibility is highly accepted and broadly described in the literature. At present, the greatest advances in the study of this genetic component have been achieved through GWAS, which analyze thousands of patients and ethnically matched controls searching for evidences to explain MS pathogenesis. However, this kind of studies does not explore genetic variants other than SNPs and the heritability underlying this and other complex diseases persists partially unexplained.LILRA3 deletion on MS risk; however, their results were not fully concordant and it was difficult to draw a firm conclusion regarding the impact of this gene on MS. Two studies performed in German , with modest evidence of heterogeneity (I2 = 25%).Previously reported studies tried to evaluate the influence of the n German  and Spann German  showed a controls, and it equency 4.1%, but  controls could noLILRA3 deletion on R-MS susceptibility by studying this association in a large number of individuals from three different Spanish locations  and performing a meta-analysis with the available data. Overall, our study with 6380 individuals  did not reveal any significant difference when the distribution of the LILRA3 deletion in MS patients and controls was compared , albeit in the context of a moderate evidence of heterogeneity (I2 = 32%). Substantial attenuation of the latter was achieved after the sensitivity analysis, which showed that the exclusion of the “2009-Spain (Madrid)” cohort reported by Ordonez et al. [LILRA3 deletion by itself does not alter MS susceptibility.Considering that small sample sizes might explain the inconsistent results observed in the different reports, we aimed to ascertain the influence of the HLA-DRB1*15:01 as the main risk allele described in MS and the putative gender bias,which might underlie the inconsistent results reported in the different studies. As previously described by Ordonez et al. [HLA-DRB1*15:01 and the LILRA3 deletion in order to confer MS risk, but no significant epistatic interaction was observed (p = 0.36). Moreover, to evaluate whether the LILRA3 deletion presented a possible gender bias in MS risk, as described by Du et al. in rheumatoid arthritis susceptibility [DRB1*15:01- detailed information. Therefore, stratifications considering either the HLA-DRB1*15:01 allele by itself or including also gender information did not evidence global association of LILRA3 deletion with MS risk, demonstrating that a more stringent stratification of patients did not explain previous results.Additionally, we also considered two other important factors: z et al. ,we assestibility , the epiLILRA3 gene conferring risk to different autoimmune diseases and this gene would be another example of the well-established concept of shared genetic background among autoimmune diseases [2 = 0.83) with the LILRA3 deletion and, therefore, it was selected by Du et al. to study the association of the LILRA3 gene with rheumatoid arthritis susceptibility[LILRA3 deletion has not been formally reported. Nonetheless, this SNP was included in an analysis of immune-related loci that identified 48 new risk variants for MS [LILRA3 gene are similar to that displayed by the LILRA3deletion, no association with either MS or rheumatoid arthritis, psoriasis, type 1 diabetes or celiac disease risk was identified in this region, in agreement with the present results.Some works seemed to support the role of the diseases , 15, 17.diseases . It is wLILRA3 deletion and the disease in a European population [LILRA3 does not seem to be underlying MS pathogenesis. Understanding the full load of genetic factors contributing to disease risk has potential for profound improvements in health care. Yet, the known genetic factors account for only a small portion of the estimated heritability of complex phenotypes and every effort to overcome this impasse should be emphasized.Our results are also concordant with those found in psoriasis, that reported lack of significant association between the pulation , althougS1 Checklist(DOCX)Click here for additional data file.S1 PRISMA Checklist(DOC)Click here for additional data file.S1 Table(DOCX)Click here for additional data file.


Obese women had a higher risk of GBC than men did , and there was significant association between overweight and GBC risk in women , but not in men . CONCLUSIONS: Findings from this meta-analysis indicate that obesity is associated with an increased risk of GBC, especially in women. Overweight is associated with GBC risk only in women.OBJECTIVES: To provide a quantitative assessment of the association between excess body weight, interpreted as increased body mass index (BMI), and the risk of gallbladder cancer (GBC). METHODS: We identified eligible studies in Medline and EMBASE up to 1 February 2015, and reference lists of retrieved articles. Summary relative risks with their 95% confidence intervals were calculated in a random-effects model. Subgroup analyses were performed according to study design, gender, geographic location, ascertainment of exposure and adjustment for confounders. RESUITS: A total of 12 cohort studies and 8 case-control studies were included in the meta-analysis. Overall, compared with “normal” weight, the summary relative risks of GBC were 1.14  for overweight individuals (BMI 25–30 kg/m Gallbladder cancer (GBC) is a highly fatal malignancy that differs from other cancers of the biliary tract, as being approximately two to five times more common in women than in men . Prognos2) or obesity (BMI > 30 kg/m2), is increasingly recognized as an important risk factor for various cancer types. Over the past decades, evidence from clinical studies has addressed the possible link between excess body weight and risk of GBC, but the findings have been somewhat contradictory. Early studies found no statistically significant results [It has been established that history of gallstone is the leading cause of gallbladder cancer worldwide . Additio results ,9,10, wh results ,12.Our clinical observations indicate a high frequency of obesity among patients with GBC. In the present study, we therefore carried out a systematic review and meta-analysis of all available evidence of observational studies following the meta-analysis of observational studies in epidemiology (MOOSE) guidelines  to clariTwo authors independently performed a literature search using Medline and EMBASE database up to 1 February 2015 with the following text words and/or Medical Subject Heading (MeSH) terms: “body mass index”, “BMI”, “overweight”, “obesity” or “excess body weight”, combined with “gallbladder cancer”, “gallbladder neoplasm” or “biliary tract cancer”. We also reviewed the reference lists of retrieved articles to search for additional studies. No language restrictions were imposed.Published articles were included according to the following criteria: (1) the outcome of interest was GBC incidence or mortality; (2) the exposure of interest was overweight or obesity defined by BMI; (3) estimates of odds ratio (OR) or relative risk (RR) with corresponding 95% confidence intervals (CIs)  were reported. Two authors independently evaluated all of the studies retrieved from the databases. Any discrepancies between the two reviewers were solved by joint reevaluation of the manuscript. If there were multiple publications from the same study, the most comprehensive one which could provide detail information for subgroup analysis was selected, using other publications to clarify methodology or characteristics of the population.versus self-reported), and the effect estimates with 95% CIs. When studies provided more than one RR, we extracted all of them and applied the data according to subgroup analysis. The quality of each study was assessed independently by three reviewers using the Newcastle-Ottawa Scale (NOS). The NOS consists of three parameters of quality: selection, comparability, and outcome (cohort studies) or exposure (case-control studies). The NOS assigns a maximum of four points for selection, a maximum of two points for comparability, and a maximum of three points for exposure or outcome [Three authors independently evaluated all of the studies retrieved according to the aforementioned inclusion criteria. Discrepancies between the three reviewers were solved by a joint reevaluation of the original article. The following information from each included study was extracted: the first author’s last name, geographic location, year of study conducted, sample size, study design, gender and age of participants, duration of follow-up (cohort studies), BMI categories, assessment of BMI (measurement  outcome . Any dis2) and obesity (BMI > 30 kg/m2 or a discharge diagnosis of obesity) compared with “normal” weight (BMI 18.5–24.9 kg/m2). If studies reported relative risk separately for men and women, we combined the gender-specific estimates to the pooled analysis. When non-standard BMI categories were provided, we selected the category that was most closed to those defined by the WHO. Summary relative risk (SRR) estimates with their corresponding 95% CIs were combined in a random-effects model. Subgroup analyses were performed according to study design (cohort and case-control studies), gender (men and women), and geographic location (non-Asia and Asia), BMI assessment (measurement and self-reported), Follow-up time (>10 years and <10 years), smoking status (smokers and non-smokers), Alcohol abuse (Yes and No). We performed sensitivity analysis to estimate the influence of each individual study on the summary results by repeating the random-effects meta-analysis after omitting one study at a time.To examine associations between overweight/obesity and the risk of GBC, we computed SRRs for two categories of BMI as defined by the World Health Organization (WHO) for adults: overweight . Finally, a total of 12 cohort studies ,24,25,26Q = 23.85, I2 = 24.9%; obesity: SRRs = 1.56, 95% CI = 1.41–1.73, Q = 37.26, I2 = 15.4%) compared to “normal weight”. We then performed subgroup analyses by study design, gender, geographic location, ascertainment of exposure and adjustment for confounders, as shown in As shown in In stratified analysis by study design, a statistically significant positive association between BMI and GBC risk was observed for the cohort studies . Moreover, in cohort studies with follow-up time >10 years, overweight and obesity were strongly associated with incidence of GBC , while only overweight was observed associated with GBC risk in cohort studies with follow-up time < 10 years . For the case-control studies, only obesity was strongly associated with GBC risk . The SRRs of GBC incidence for obesity in population-based case-control studies was 1.43 (95% CI = 1.09–1.89); no significant association between obesity and GBC risk was observed in hospital-based case-control studies.A significant gender-specific difference was observed in the association between obesity and GBC risk, and obese women had a higher risk of GBC . However, overweight men are not associated with risk of GBC .In stratified analysis by geographic location, the association between obesity and the risk of GBC was similar for both Asia and non-Asia . For nonIn addition, when stratified by potential confounders, overweight people with smoking and alcohol consumption were strongly associated GBC risk, no significant association between overweight and the risk of GBC was found in those without smoking and alcohol consumption , indicating that smoking and alcohol consumption are positive confounders. No differences were observed in the association between obesity and GBC incidence when stratified by smoking and alcohol consumption.et al. [I2 = 17.8%).In the sensitivity analyses, we removed one study at a time and calculated the SRRs. We found that there were no changes in the direction of effect when any one study was excluded, supporting the robustness of our results. For example, when the study of Engeland et al.  was exclp = 0.483) results  This meta-analysis was based on 20 epidemiologic studies, which might minimize the possibility of selection bias. (2) Most of the included studies provided more than one RRs, which could be applicable to accurately subgroup analysis. (3) The included studies evaluated multiple confounders including smoking and alcohol. The relationships between BMI and risk of GBC in each study were derived from regression after adjustment at least for age and gender.Our meta-analysis has limitations that affect interpretation of the true results. First, inadequate control for confounders may bias the results, leading to exaggeration or underestimation of risk estimates. Thus, when interpreting the association between excess body weight and GBC risk, possible unmeasured or residual confounding factors should be considered. Five studies were of acceptable quality (NOS < 7), mainly due to the adjustments made for confounders. Smoking and alcohol abuse is closely related to GBC risk. Subgroup analysis results also shown that overweight people with smoking and alcohol consumption were strongly associated GBC risk, while no significant association between overweight and the risk of GBC was found in those without smoking and alcohol consumption, suggesting that data from unadjusted studies might lead to an overestimation of overweight in the development of GBC. Interestingly, no differences were observed in the association between obesity and GBC incidence when stratified by smoking and alcohol consumption, suggesting that obesity might be an independent risk factor. Gallstone is closely related to GBC risk . MeanwhiI2 = 31.9%, so we could combine studies in a meta-analysis. We analyzed this review in both fixed effects and random effects, and found that they had no significant differences. Thus, the more conservative one, random effects, was chosen finally. Next, when we tried to carry out subgroup analysis to investigate sources of heterogeneity, statistical heterogeneity was lower in analysis of case-control studies, population based studies, Non-Asia studies and studies of BMI ascertainment by self-report, indicating that these might account for heterogeneity observed in studies about overweight and GBC risk.There was significant heterogeneity observed across studies about overweight and GBC risk, but the heterogeneity is low and acceptable with In summary, findings of this meta-analysis provide evidence that excess body weight may increase GBC risk. Further studies that meet strict criteria on this subject are needed to strengthen the association between BMI and GBC risk, especially those adjusting potential confounding factors such as gallstones and DM.


There was evidence of an increased risk of CHD in the highest compared with the lowest categories of the unhealthy/Western-type dietary patterns . The results of this meta-analysis indicate that different dietary patterns may be associated with the risk of CHD.Previous studies reported the potential associations between dietary patterns and the risk of coronary heart disease (CHD) in adulthood, however a consistent perspective has not been established to date. Herein, we carried out this meta-analysis to evaluate the associations between dietary patterns and the risk of CHD. MEDLINE and EBSCO were searched for relevant articles published up to April 2015. A total of 35 articles  met the inclusion criteria and were included in the present meta-analysis. The decreased risk of CHD was shown for the highest compared with the lowest categories of healthy/prudent dietary patterns (odds ratio (OR) = 0.67; 95% confidence interval (CI): 0.60, 0.75;  Although the incidence and mortality of coronary heart disease (CHD) have decreased in the United States and Western Europe since 1970s, it remains the leading cause of death globally, with 7.2 million deaths occurring worldwide every year . In ChinIn the past few decades, many studies particularly focused on diet modification as an important determinant in the development of CHD and found associations between the intakes of individual foods or nutrients and the risk of CHD. However, in reality, people generally do not take nutrients alone but consume meals containing many combinations of foods and nutrients . ConsequRecently, an emerging body of evidence has suggested that there has been considerable attention in epidemiological research on the associations between overall dietary patterns and the risk of CHD ,21,22,23An electronic literature search was conducted in MEDLINE and EBSCO to identify human studies written in English or Chinese, published up to April 2015, with the following keywords: dietary pattern, dietary patterns, alcohol drinking, alcohol consumption, cardiovascular disease, ischemic heart disease, myocardial infarction, coronary heart disease and coronary diseases. In addition, we manually searched all references cited in original studies and reviews identified.Two independent reviewers (L. Shu and X.-Y. Zhang) read the abstracts of the papers retrieved in the initial search to identify studies that examined the associations between dietary patterns and the risk of CHD. When all reviewers agreed, the papers were reviewed against inclusion and exclusion criteria for this meta-analysis. To be eligible, dietary pattern studies had to fulfill the following criteria: (1) The study was published as an original article reporting the relationship of different dietary patterns and the risk of CHD. (2) Food or dietary patterns in studies were examined by principal component analysis (PCA) and/or factor analysis (FA). (3) Odds ratios or hazard ratios, and percentage for CHD  had been provided. (4) CHD was diagnosed based on clinical manifestations , electrocardiogram, and coronary aeteriography. Alcohol studies were included in this meta-analysis if they met the following criteria: (1) the study was published as an original article; (2) the association between alcohol intake and risk of CHD had been reported in studies; (3) the different alcohol consumption categories have been described in studies; and (4) CHD was diagnosed based on clinical manifestations, electrocardiogram, and coronary arteriography.Information extracted from each study included authors, geographic region, study design, sample size, the number of CHD, dietary assessment method, identification of dietary patterns, and factors that were adjusted in the including studies.Dietary patterns were identified by principal component analysis or factor analysis. Factor scores for each pattern were categorized into tertiles, quartiles, or quintiles  The different forms of alcohol consumption were converted into grams of ethanol per day. Alcohol consumption of >25 g/day for men or >12.5 g/day for women was defined as a high intake of alcohol or heavy alcohol drinking; alcohol consumption of <12.5 g/day for men or <7.5 g/day for women was defined as a low intake of alcohol, and alcohol consumption of >12.5 g/day and <25 g/day for men or >7.5 g/day and <12.5 g/day for women was defined as a moderate alcohol consumption .i.e., with a score of six or higher) were considered to be of high methodological quality.The Newcastle–Ottawa Quality Assessment scale was used for quality assessment . Eight qp Values less than 0.05 were considered to be significant. In this meta-analysis, a random-effects model was used to account for possible heterogeneity between studies, while a fixed-effects model was adopted in the absence of heterogeneity [Heterogeneity of the study results was estimated by the chi-squared test. ogeneity .versus the lowest categories of healthy/prudent, unhealthy/Western-type dietary patterns and alcohol consumption. Random-effect models were used to calculate the pooled odd ratio (OR) for dietary patterns in highest categories compared with lowest categories. Raw data from individual studies were weighted and combined to produce an overall OR. Publication bias was assessed by inspection of the funnel plot and by formal testing for “funnel plot” asymmetry using Begg’s test and Egger’s test [p values less than 0.05 were considered significant.Statistical analyses were performed using Review Manager, version 5.0  and STATA software, version 12.0 . The original studies reported the results of dietary patterns or alcohol consumption in terms of quintiles, quartiles, and tertiles of dietary factor scores and the risk of CHD. We used meta-analysis to evaluate the risk of CHD in the highest r’s test . All stan = 173); title and abstract did not contain the data on classification of dietary pattern or alcohol consumption (n = 278); did not provide sufficient dichotomous data on dietary pattern and CHD (n = 25); focused on single nutrients or food (n = 18); did not provide the data about percent of CHD or number of each group (n = 20); reported data using different alcohol consumption categories (n = 10); and reported the same data (n = 1). At last, 35 articles [An electronic literature search in the database of MEDLINE and EBSCO identified 560 studies, 525 of which were excluded based on the reasons listed in on n = 27; did notfood n = ; did notroup n = ; reportearticles ,47,48,49p < 0.00001). A random-effects model was used to assess the data included in our analyses. The heterogeneity was apparent in all the studies .The healthy/prudent dietary patterns were characterized by high consumption of vegetables, fruits, whole grains, olive oil, fish, soy, poultry and low fat dairy. The examined studies labeled it as “Prudent” ,13,15,23p = 0.02) where all studies were combined in the random-effects model. There was significant heterogeneity .The unhealthy/Western-type dietary patterns were characterized by high consumption of red and/or processed meat, refined grains, sweets, high-fat dairy products, butter, potatoes and high-fat gravy, and low intakes of fruits and vegetables. The studies under consideration labeled it as “Western” ,15,23,28I2 = 83%, p < 0.00001). p < 0.00001).The alcohol consumption was characterized by moderate intakes of wines, alcohol-containing beers, and white spirits. Nineteen studies  have identified moderate alcohol consumption. The data from nineteen studies were assessed using the random-effects model, and there was significant heterogeneity .Inspection of funnel plots did not reveal evidence of asymmetry . Begg’s The quality of each study in terms of population and sampling methods, description of exposure and outcomes, and statistical adjustment of data, is summarized in p < 0.01; alcohol consumption: OR = 0.68; 95% CI :0.51, 0.80; p < 0.05). When sample size <10,000 was removed [p < 0.01). Similarly, when the results were analyzed by removing the studies adjusted for total energy intake [p < 0.01). As these variables have a strong effect on association between different dietary patterns and risk of CHD, their differences may partially explain the observed heterogeneity between studies  [Limited epidemiological research has reported the associations between dietary patterns and the risk of CHD. To our knowledge, this is the latest meta-analysis evaluating the evidence for dietary patterns and CHD risk. In the present study, we have an update on the earlier systematic review  ,51 and fet al. reported an inverse association between dietary fiber and the risk of CHD [In our analyses, the healthy/prudent dietary patterns were associated with a reduced risk of CHD. Our results were in agreement with some previous studies, which reported an inverse association between healthy/prudent dietary patterns and the risk of CHD ,23. The k of CHD . To our k of CHD . In addik of CHD ,58.The unhealthy/Western-type dietary patterns were associated with an increased risk of CHD in this meta-analysis. Our results were consistent with previous studies ,19, whicAn inverse association was shown for moderate alcohol consumption and the risk of CHD in our analyses. Previously, a meta-analysis of alcohol consumption and the risk of CHD indicated that moderate alcohol-drinking (≤1.5 drinks/day) was associated with a decreased risk of CHD . In factet al. and Hou et al., 2014) [This meta-analysis has its own strengths and limitations. Firstly, this is the latest meta-analysis reporting the associations between dietary patterns and the risk of CHD. We not only have an update on an earlier systematic review  ,51, but In conclusion, results from this meta-analysis indicate that healthy/prudent dietary patterns and moderate alcohol consumption are associated with a decreased risk of CHD, while unhealthy/Western-type patterns are associated with an increased risk of CHD. In addition, these findings also suggest that a change of diet is essential for the prevention of CHD. Herein, it makes sense to elucidate the potential associations between dietary patterns and the risk of CHD, and provide scientific rationale for formulating dietary guidelines. Further studies are required to confirm the causal associations between dietary patterns and the risk of CHD.


NBS1 participates in detecting the DSBs and plays an essential role in maintaining genomic stability. Single nucleotide polymorphisms (SNPs) in NBS1 gene were commonly tested that associated with the susceptibility to multiple cancers, but the results remained controversial. Thus, we conducted two independent hospital-based case–control studies comprising 1,072 colorectal cancer patients and 1,263 controls to evaluate the association between four NBS1 SNPs and colorectal cancer risk. The result showed that rs2735383C/G polymorphism in the 3’-untranslated region (UTR) of NBS1 was significantly associated with risk of colorectal cancer using logistic regression (P<10-4). Furthermore, we observed that rs2735383CC genotype was associated with substantially increased risk of colorectal cancer , compared with the rs2735383GC+GG genotypes. Further functional experiments demonstrated that the rs2735383C allele in the NBS1 disrupted the binding affinity of has-miR-509-5p to the NBS1 3’-UTR in colorectal cancer cells, affecting the NBS1 transcriptional activity and expression level. In conclusion, current evidence suggests that the rs2735383C/G polymorphism might contribute to the risk for colorectal cancer.As the central protein of the double strand breaks (DSB)-induced DNA repair pathway,  Colorectal cancer (CRC) is one of the most common malignancies leading cause of cancer-related deaths throughout the world. In the NBS1 plays essential roles in the assembly of MRN complexes and directs MRN complexes coordinate DSBs to processing and repair [NBS1 gene may modulate DNA repair capacity, and affect the propensity to cause cancer. Accumulating studies have explored the associations between common polymorphisms in NBS1 and human cancers risk [NBS1 may influence the binding ability of microRNAs, thus may affect gene’s expression [NBS1 polymorphisms and risk of CRC in Chinese population. Therefore, we hypothesized that common polymorphisms of the NBS1 gene may associate with the risk of CRC.DNA repair pathways play a key role in maintaining genomic integrity and preventing cells against cancer. In contd repair , 18. Theers risk –22. For ers risk . Furtherpression . HoweverBased on these hypotheses, we performed two hospital-based case-control studies to evaluate the possible influence of common polymorphisms  on CRC risk in Chinese population.All individuals were genetically unrelated ethnically homogenous Han Chinese, and eligible participants were 21–90 years of age at recruitment during the study period. For the discovery set, a total of 763 patients diagnosed CRC and 892 controls frequency-matched to the cases with regards to age (±5 years), sex and residential area were consecutively recruited from the People’s Hospital of Yichun City  and China-Japan Friendship Hospital . For the validation, 313 CRC patients and 371 controls frequency-matched to the cases with regards to age (±2 years), sex and residential area were consecutively enrolled from the China-Japan Friendship Hospital . Among eligible participants, the recruitment rates were about 95% for patients and about 81% for controls. All patients hadn’t received any chemotherapy or radiotherapy and there were no age, stage, or histology restrictions. Tumor staging and pathological type were evaluated according to the 2002 American Joint Committee on Cancer staging system. The participation rate for controls was 89%. The clinical characteristics of all patients and controls of the discovery set and validation set are listed in NBS1 SNPs in the genome context, four NBS1 SNPs  previously reported in relation to disease outcomes including cancers were selected and analyzed [NBS1 SNPs were genotyped by allele-specific MALDI-TOF mass spectrometry  as described before[Based on the possibility of functional change of analyzed , 25. Gened before, 27 within vitro. The full-length of human NBS1 3’-UTR flanking SNP rs2735383 polymorphism was synthesized by the Genewiz Company and were then cloned into the psi-CHECK2 basic vector . The result revealed that the conversion from C to G of the 3’-UTR rs2735383 polymorphism cause new binding site of microRNA hsa-miR-629, has-miR-499-5p, and has-miR-509-5p. For the transient transfection assay, human colorectal cell lines, namely HCT116 and HT-29 were seeded at 1×105 cells per well in 24-well plates . The luciferase assays were performed as described previously [NBS1-3’-UTR vectors were cotransfected with 40pmol microRNA mimics  and with or without 40pmol inhibitors by Lipofectamine 2000  according to the manufacturer’s instructions. After transfection for 24h, cells were collected and renilla luciferase activity was measured with the Dual-Luciferase Reporter Assay System  (Promega). Three independent experiments were done for each plasmid construct.Because the essential role of 3’-UTR in gene, we predicted the rs2735383C/G polymorphism in the 3’-UTR by our bioinformatics analysis  and reversed transcribed to cDNA using oligo-dT primer. We carried out ABI Prism 7500 sequence detection system to evaluate NBS1 and GAPDH mRNA levels based on the SYBR-Green method. The latter was used as an internal quantitative control and NBS1 expression level was normalized to GAPDH. The TaqMan MicroRNA Assays (Applied Biosystems) was used to detect hsa-miR-509-5p expression according to the manufacturers’ protocol and the universally expressed U6 small nuclear RNA was used for an internal control.29 CRC tissues with different rs2735383 genotypes were subjected to investigate the correlation between the NBS1 genotypes. One-way ANOVA were used to evaluate the differences in luciferase reporter activity and the effect of rs2735383C/G on the NBS1 expression among the CRC patients. The statistical power was calculated by using the PS Software (http://biostat.mc.vanderbilt.edu/twiki/bin/view/Main/PowerSampleSize). All tests were two-sided by using the SAS software . Statistically significant level was defined as P<0.05.The differences in the frequency distribution of selected demographic variables as well as rs2735383C/G allele and genotypes between cases and controls were assessed by the chi-square tests. The Hardy–Weinberg equilibrium (HWE) of the cancer-free controls genotype distributions was tested by a goodness-of-fit chi-square test. The strength of the association between the rs2735383C/G polymorphism and CRC cancer risk was estimated by odds ratios (ORs) with 95% confidence intervals (CIs) using unconditional logistic regression and adjusted for age and sex. Group’s comparison was analyzed by the Student’s t-test (two-sided). Stratified analysis was also performed by subgroups of age, sex or other characteristics  to evaluate the stratum variable-related ORs among the NBS1 gene in 1076 CRC patients and 1263 healthy controls. The genotype frequencies of these polymorphisms in controls were consistent with the HWE . The genotype distributions of NBS1 polymorphisms of the cases and the controls in the discovery and validation set are presented in P<10−4 for rs2735383C/G in the discovery set and P = 0.039 for rs2735383C/G in the validation set). As showed in P<10−4 for the discovery set and adjusted OR = 1.60; 95% CI = 1.11–2.37, P = 0.025 for the validation set). Furthermore, significant association was observed between rs2735383 polymorphisms and CRC risk when combined the two populations (P<10−4).In the study, we genotyped four common polymorphisms  of P = 0.004). However, there was no statistically difference between the distributions of other selected variables and CRC risk.We further performed the stratification analyses to evaluate the effects of rs2735383C/G genotypes on CRC risk according to age, sex, smoking status, alcohol consumption and family history of cancer in the discovery set. As showed in NBS1 expression in CRC cells. The result showed that CRC cells transiently co-transfected hsa-miR-509 mimics and vectors containing the rs2735383C allele significantly decreased luciferase activity, compared with cells vectors containing the rs2735383G allele  than patients with the GC (0.363±0.047) or GG  genotypes. Additionally, QRT-PCR was further used to examine whether hsa-miR-509-5p is constitutively expressed in CRC tissue samples. As shown in NBS1 rs2735383C/G genotypes in tumor tissues collected from CRC patients (P = 0.345).We further examined the NBS1 SNPs in the susceptibility to CRC in Chinese populations. We found that hsa-miR-509-5p might bind tightly to 3’-UTR of NBS1 containing the rs2735383C allele, negatively regulating the level of NBS1. Furthermore, in the following stratified analyses, it appeared that a high risk effect of this polymorphism was especially evident among subgroups of drinkers.CRC carcinogenesis is multifactorial. Not only environmental risk factors but also genetic factors might play an essential role in the etiology of CRC. In this hospital-based case-control study, we investigated the role of four common NBS1 is a key regulator that participates in DSB repair by directing the MRN protein complex to the sites of DNA damage and stimulating its DNA binding and nuclease activity. The main functional domains of NBS1 in DNA damage responses comprising the conserved forkhead-associated (FHA) domain and the breast cancer carboxy-terminal (BRCT) domain, which are important in the recognition and repair of aberrant DNA [NBS1, an important part of MRN complex, in maintaining genomic stability and preventing the tumorigenic processes by recruiting repair and checkpoint proteins during the DNA breaks[NBS1 may affect the function of NBS1 and protein-protein interaction, and further probably increase susceptibility of cancer. As one of the most commonly investigated polymorphisms, rs1805794G/C is located in the BRCT domain and influence the formation of a BRCA1-associated genome surveillance complex (BASC) though binding to BRCA1, which is responsible for detecting and repair of aberrant DNA structures[NBS1 rs1805794G/C variant genotypes and cancer risk is not accordant in different cancers and ethnic group. For an example, P. Widlak found that rs1805794 variant genotypes were not associated with susceptibility of CRC in a Polish population [rant DNA –31. NumeNA breaks, 33. Thutructures. Severaltructures, 34–37. NBS1 31129G/A and 924T/C included in our study have rarely been investigated in human tumors [NBS1 gene. Several studies have investigated the rs2735383C/G polymorphism with cancer susceptibility. A meta-analysis based on thirteen eligible studies including 7561 cases and 8432 control subjects investigated that no significant association was found between the polymorphisms in NBS1 gene and bladder cancer[NBS1 in vitro, which is consistent with the previous findings [NBS1 resulting from the genetic variants in NBS1 consequently affect the NBS1 protein and other repaired proteins interact in DNA repair processes, thus may modulate the CRC susceptibility.To our knowledge, n tumors , 25. Ander cancer, breast er cancer, lymphoier cancer. Recentler cancer. In our NBS1 SNPs and the risk of CRC. We had a fairly large sample size for the case-control study. Moreover, the fact that the genotype frequencies among controls fit the Hardy–Weinberg disequilibrium law suggested the randomness of subject selection. And we have obtained a 91% study power  to detect an OR of 1.52 for the rs2735383CC genotypes (which occurred at a frequency of 16.14% in the controls) compared with the rs2735383GC+GG genotypes in the discovery set, which may have enhanced the noteworthy of our finding.This is the first study that we comprehensively investigated the associations between NBS1 gene may be a genetic susceptibility factor for the risk of CRC in Chinese population. Moreover, our findings would require larger case–control studies and well-designed mechanistic studies to verify.In conclusion, our study suggests that polymorphism rs2735383C/G in the


The fellow eye of patients with unilateral neovascular age-related degeneration (nAMD) is at increased risk of developing late AMD. Several cohort studies have evaluated the prevalence of pseudodrusen and the association between pseudodrusen and late AMD in the fellow eye of patients with unilateral nAMD. However, these studies have limited sample sizes and their results are inconsistent.To evaluate the prevalence rate of pseudodrusen, and the association between pseudodrusen and incidence of late AMD (nAMD and geographic atrophy (GA)) in the fellow eye of patients with unilateral nAMD.The PubMed, EMBASE, Web of Science, and Cochrane Library databases were searched up to July 2015, as well as other systematic reviews.All cohort studies for pseudodrusen with late AMD in the fellow eye of patients with unilateral nAMD.2 test.The numbers of patients with and without pseudodrusen at baseline and the numbers of incident nAMD and GA during follow up among patients with and without pseudodrusen were independently extracted by 2 authors. The results were pooled using random-effects meta-analysis. Heterogeneity was assessed using the IPrevalence rate of pseudodrusen, risk ratios (RRs) and their 95% confidence intervals (95% CIs) for associations between pseudodrusen and the incidence of nAMD and GA in the fellow eye.2 = 87%). Pseudodrusen were associated with an increased risk of nAMD , GA , and late AMD .Five cohort studies (N = 677 patients) from 8 countries across 4 continents were included. The pooled prevalence rate of pseudodrusen in the fellow eye was 48.1% (95% Cl: 36.7–59.5%, IFor patients with unilateral nAMD, pseudodrusen were present in about half of the fellow eyes. The presence of pseudodrusen was associated with a 1.5 times higher risk of developing nAMD, a 4.7 times higher risk of developing GA, and a 2 times higher risk of developing late AMD. Pseudodrusen should be considered in evaluating the risk of late AMD development; however, due to considerable heterogeneity across these studies, a larger study is needed to validate these findings. Age-related macular degeneration (AMD) is the leading cause of blindness among the elderly in Western countries and is the third leading cause of blindness in the world., 2 Many The Beaver Dam Eye Study and the Blue Mountains Eye Study reported 4.0% and 6.6% 15-year pseudodrusen incidence rates based on the grading of standard color fundus photography and found that the proportion of eyes with pseudodrusen that progressed to late AMD within 5 years was four-fold to six-fold higher than eyes without pseudodrusen but with other early AMD features. Several This meta-analysis was conducted following the Meta-analysis of Observational Studies in Epidemiology (MOOSE) statement.The PubMed, EMBASE, Web of Science, and Cochrane Library databases were searched up to July 2015 using combinations of the following search terms:  AND fellow eye. No language restrictions were imposed.The titles and abstracts of the retrieved articles were reviewed to exclude any clearly irrelevant studies. The full texts of the remaining articles were then reviewed to determine whether they contained any information on the prevalence of pseudodrusen and the association with the incidence of late AMD (either nAMD or GA). To find all relevant studies, the reference lists of systematic reviews on pseudodrusen,–25 as weThe inclusion criteria for this meta-analysis were: (1) the study patients had unilateral nAMD; (2) the presence of pseudodrusen was evaluated in the fellow eye using at least two imaging modalities; and (3) the study outcomes were incidence of either nAMD or GA. We (QZ and GSY) independently assessed the eligibility of studies and extracted data from each eligible study, and the discrepancies were discussed to reach agreement.The study characteristics and results from eligible studies were extracted, including the name of the first author, the year of publication, the sample size, the country of study, the mean age of participants, the number of years of follow-up, the imaging modalities used for detecting pseudodrusen, the prevalence of pseudodrusen at baseline, and the incidence of nAMD and GA. We contacted the corresponding authors of retrieved articles if additional data were needed.http://www.cebm.brown.edu/open_meta/download.html) was used for the statistical analysis. Random-effects models were used to estimate the pooled prevalence rate of pseudodrusen and to estimate the risk ratios (RRs) and their 95% confidence intervals (95% CIs) for the associations of pseudodrusen with the incidence of nAMD, GA and late AMD. The pooled RR and its 95% CI for CNV and GA were not adjusted by other covariates, because the covariates-adjusted RRs for CNV and GA were not reported in most studies. However, for late AMD, the pooled covariate-adjusted RR and its 95% CI were calculated from the individual covariates-adjusted RR  from 3 studies.[2 statistic.[2 ranges between 0% and 100% (0% represents no heterogeneity), and higher values represent larger heterogeneity. To account for the various length of follow-up across studies, we also calculated the annual incidence rate of nAMD, GA and late AMD by using the person years method. A sensitivity analysis, removing each study one at a time, was performed to confirm the robustness of our findings.Open Meta-Analyst software . Correspondingly, the annual incidence of nAMD in patients without pseudodrusen ranged from 6.5% to 26.8%, with a pooled annual incidence rate of 12.9% (95% Cl: 6.7%-19.1%). In the pooled analysis, nAMD occurred in 122 (39.5%) of 309 patients with pseudodrusen and in 89 (24.2%) of 368 patients without pseudodrusen, for an RR of 1.54 .Among the 3 studies that reported the incidence of both nAMD and GA,–29 the aTo evaluate the robustness of the above results, we performed a sensitivity analysis by removing one study at a time and re-calculating the prevalence rate of pseudodrusen and the RRs for nAMD, GA, and late AMD. In these sensitivity analyses, the pooled estimates of the prevalence rate of pseudodrusen ranged from 46% to 52%, indicating a robust estimate of the pseudodrusen prevalence rate. Similarly, the pooled RR for the nAMD ranged from 1.5 to 1.7, indicating the result is not substantially influenced by any single study. However, the estimate of RR for GA was substantially influenced by the study by Pumariega et al. in that the pooled RR is 10.4 when this study is excluded, and 3.4 to 3.8 when this study is included. The pooled RR for late AMD is substantially influenced by the study of Hogg et al. in that the pooled RR for late AMD is 1.9 when this study is excluded, and approximately 3.6 when this study is included.This meta-analysis evaluated the prevalence rate of pseudodrusen and its association with the incidence of nAMD and GA in the fellow eye of patients with unilateral nAMD. Our meta-analysis found that the prevalence rate of pseudodrusen in the fellow eye was high (48.1%) and that pseudodrusen was associated with an increased risk of late AMD development, with a 1.5 times higher risk of nAMD, a 4.7 times higher risk of GA, and an approximately 2 times higher risk of late AMD in the fellow eye with pseudodrusen compared to fellow eyes without pseudodrusen.The association of pseudodrusen with nAMD and GA had similar magnitude of risk ratio as the other risk factors of late AMD such as age, smoking, large drusen and pigmentation changes.–35 ConsiThe prevalence rate of pseudodrusen in the fellow eye in the included studies varied from 34.7% to 58.0%. The earliest study, which pIn contrast to the detection of pseudodrusen, the determination of nAMD in the individual studies was based on fluorescein angiography and/or indocyanine green angiography. The incidence of nAMD depended on the length of follow-up. In these studies, the mean length of follow-up ranged from 2 to 4 years. To account for the differential length of follow-up, we calculated the annual incidence rate of late AMD using person year analyses. The annual incidence rate of nAMD in the fellow eyes of patients with pseudodrusen was higher than the fellow eyes without pseudodrusen (15.3% vs. 12.9%). A previous meta-analysis found the cumulative incidence rate of nAMD in the fellow eye to be 11.1% to 12.2% after 1 or 2 years, respectively[Five studies evaluated the association of pseudodrusen with the incidence of nAMD in the fellow eye and the reported results are quite inconsistent, with RRs ranging from 0.82 to 5.77. Our meta-analysis of these 5 studies found that pseudodrusen are statistically significantly associated with nAMD with a RR of 1.54 (95% CI: 1.10–2.16), demonstrating that pseudodrusen conferred an increased risk of nAMD development in the fellow eye of patients with unilateral nAMD. This result suggests that even among eyes at a high risk of developing AMD, the presence of pseudodrusen may exacerbate the development of nAMD. This association may be related to a thinning of the choroidal thickness in eyes with pseudodrusen,–43 which2 = 64%). It is uncertain whether the adjustment of well-known risk factors of GA  will substantially change the association because these covariates-adjusted risk ratios from individual studies are not available for calculating the pooled covariate-adjusted risk ratio for GA. Future large cohort studies are needed to validate this association. Similar to the findings from this meta-analysis, cross-sectional studies have also found that the presence of pseudodrusen was associated with the presence of GA in AMD patients.[Three studies evaluated the association between the presence of pseudodrusen and the incidence of GA in the fellow eye of patients with unilateral neovascular AMD.–29 The rpatients.–46 Althopatients. but the patients. The prespatients. thus conAmong the 3 studies that evaluated both nAMD and GA,–29 the pThe strength of this meta-analysis is generalizability. Our comprehensive review resulted in 5 eligible studies representing 8 countries from 4 different continents. However, the results of these studies showed a substantial amount of heterogeneity. Differences in the imaging modalities used to detect pseudodrusen, sample sizes, and lengths of follow-up likely contributed to this heterogeneity. This meta-analysis is also limited in that the pooled covariates-adjusted risk ratio for nAMD and GA could not be calculated, due to the limited covariates data from these studies. As a result, larger cohort studies are needed to validate the findings from this meta-analysis, particularly for the associations of pseudodrusen with nAMD and GA.Pseudodrusen are present in about half of the fellow eyes of patients with unilateral nAMD. The presence of pseudodrusen in the fellow eye is associated with a 1.5 times higher risk of developing nAMD, a 4.7 times higher risk of developing GA, and a 2 times higher risk of developing late AMD. Pseudodrusen should be considered as a risk factor for late AMD development.S1 File(DOC)Click here for additional data file.S2 File(DOCX)Click here for additional data file.S3 File(PDF)Click here for additional data file.


Studies concerning the association between circulating resistin and mortality risk have reported, so far, conflicting results.To investigate the association between resistin and both all-cause and cardiovascular (CV) mortality risk by 1) analyzing data from the Gargano Heart Study (GHS) prospective design ; 2) performing meta-analyses of all published studies addressing the above mentioned associations.MEDLINE and Web of Science search of studies reporting hazard ratios (HR) of circulating resistin for all-cause or CV mortality.Performed independently by two investigators, using a standardized data extraction sheet.In GHS, adjusted HRs per one standard deviation (SD) increment in resistin concentration were 1.28 (95% CI: 1.07-1.54) and 1.32 (95% CI: 1.06-1.64) for all-cause and CV mortality, respectively. The meta-analyses included 7 studies  for all-cause mortality and 6 studies  for CV mortality. Pooled HRs per one SD increment in resistin levels were 1.21  and 1.05  for all-cause and CV mortality, respectively. At meta-regression analyses, study mean age explained 9.9% of all-cause mortality studies heterogeneity. After adjusting for age, HR for all-cause mortality was 1.24 (95% CI: 1.06-1.45).Our results provide evidence for an association between circulating resistin and mortality risk among high-risk patients as are those with diabetes and coronary artery disease. Several studies indicate that resistin, a 12.5 kDa cysteine-rich pro-inflammatory adipokine, is a major promoter of atherosclerosis and related cardiovascular (CV) disease , includiSince CV disease is the first cause of mortality worldwide, resistin becomes a natural candidate to investigate as a possible mortality risk factor. In fact, conflicting results have been reported on this subject –16.To deeper understand the role of resistin on mortality rate, we firstly added new evidences by trying to replicate our previous finding  in a secThis study comprises 368 patients with type 2 diabetes mellitus (T2DM) (ADA 2003 criteria) and coronary artery disease (CAD) who were consecutively recruited at the Endocrine Unit of IRCCS “Casa Sollievo della Sofferenza” in San Giovanni Rotondo  from 2001 to 2008, as recently described , 17. AllClinical data were obtained from a standardized interview and examination. Smoking habits and history of hypertension, dyslipidemia and MI as well as glucose-lowering treatment were also recorded at time of examination. Data regarding medications were confirmed by review of medical records.Serum resistin was measured in 359 (98%) participants by a commercial ELISA  at the Research Unit of Diabetes and Endocrine Diseases in San Giovanni Rotondo, as previously described .The study was approved by the Institutional Ethic Committee IRCCS “Casa Sollievo della Sofferenza”, San Giovanni Rotondo. All participants gave written consent.The present meta-analysis follows the Preferred Reporting Items for Systematic Review and Meta-analyses (PRISMA) statements . Risks were reported as HRs along with 95% CI.Dose-response meta-analysis followed a two-step process and was performed for all-cause and CV mortalities outcomes, separately.Step 1: We considered those studies in which HRs were reported with respect to a study-specific categorical reference dose of serum resistin concentrations (i.e. tertiles or quartiles) , 11–13: Step 2: We considered those studies in which HRs were reported with respect to one increment of resistin SD  , 14–16 aStep 3: Given that resistin measurement was carried out in different specimens (i.e. serum or plasma), by different laboratories using different assays, we firstly converted the scaled measures with respect to one increment of study-specific SD, and then all study-specific slopes obtained from Step 1 and Step 2 were combined and their summary statistics were estimated. Statistical heterogeneity among studies was assessed using the Cochran Q-test and heterogeneity hold for p-values less than 0.10 . Study-sTo evaluate consistency and stability of HRs estimates found from meta-analysis and meta-regression, sensitivity analyses were carried out, by re-running all analyses after excluding one study at one time.Due to the small number of studies (fewer than ten) and the presence of substantial between-study heterogeneity, following Sterne et al.  and IoanForest plots were shown, where a square was plotted for each study whose center projection on the underlying scale corresponded to the study-specific HR. The area of the square was proportional to the inverse of the variance of the logHR and thus gives a measure of the amount of statistical information available from that particular estimate. A diamond was used to plot the summary HRs, the center of which represents the HR; the extremes of the summary HRs show the 95% CI.Bubble plot of study-specific logHRs is shown against the chosen study-level covariate which explained the most proportion of heterogeneity. The size of each bubble was inversely proportional to the standard error of specific logHR.Two-sided p-values<0.05 were considered for statistical significance. Statistical analysis were performed using SAS Software, Release 9.3  and R 2.15 (package: metafor).Clinical features of study participants are summarized in During follow-up , 81 all-cause and 58 CV deaths occurred, corresponding to an annual incidence rate of 4.2 and 3.0 events per 100 person-years, respectively.-5 and HR = 1.34, 95%CI: 1.14–1.57, p = 3x10-4, respectively; Each SD (i.e. 6.6 ng/ml) increment of serum resistin was associated with an increased risk of both all-cause and CV mortality . Sensitivity and meta-regression analyses were then performed . At sensitivity analysis, HR estimates per each unitary increment of SD in resistin levels turned out to be stable (ranging from HR = 1.13 to HR = 1.26) and quite consistent with the HR obtained from all studies. At meta-regression analysis, among all study-level covariates which were available from all studies , study mThe number of subjects from each study ranged from 150 to 1888 with a mean follow-up ranging from 1.0 to 5.5 years .With the only exception of our current study, none of the previous ones reported a significant association with CV mortality , 15, 16,The pooled HR per one increment in SD of resistin concentration for CV mortality was 1.05 , using the fixed-effects model .No significant heterogeneity across study results was observed (Q-test p for heterogeneity = 0.20), so that no meta-regression analysis was carried out.Data obtained from the GHS-prospective design confirm our previous observation that serum resistin is a predictor of all-cause mortality  and exteSince heterogeneity was observed across studies for all-cause mortality, meta-regression analysis was employed to get more reliable pooled HRs estimates. Although the observed results have to be taken with great caution because of the limited number of available studies, heterogeneity was partly explained by study mean age (i.e. stronger association in studies with older subjects).Though our meta-regression allows speculate that age acts as a modulator of the resistin effect on all-cause mortality, only studies with individual patient data may say a last word about this hypothesis. Unfortunately, no studies addressing this specific issue are available.Finally, we observed a potential risk for publication bias; however, given the small number of studies (fewer than ten) and the presence of between-study heterogeneity, we believed unnecessary to address it in accordance with previous recommendations , 24.We have to recognize some study limitations, as follows. Our meta-analyses have been conducted in high-risk patients ; so, it is unknown whether our finding applies also to “healthy” individuals with low risk profile. In addition, most studies included in the meta-analyses were carried out in patients from European Countries, thus making impossible to generalize our findings to population of different ancestry and/or environmental background.In conclusion, our results provide evidence for an association between higher circulating resistin levels and increased mortality risk. Further studied are needed to verify whether or not including resistin in risk engines which are already validated –28 improS1 PRISMA Checklist(PDF)Click here for additional data file.S1 Supporting Information(DOCX)Click here for additional data file.