Recent studies have linked certain single nucleotide polymorphisms in the leucine-rich repeat kinase 2 (LRRK2) gene with Parkinson’s disease (PD). The R1628P variant of LRRK2 may be a specific risk factor for PD in ethnic Han-Chinese populations. This study is to elucidate the epidemiological feature of R1628P in ethnic Han-Chinese population with PD. A comprehensive meta-analysis was performed to evaluate the precise association between R1628P variant and the risk for PD in ethnic Han-Chinese and subgroups stratified by gender, onset age, or family history. The analysis assessing the role of R1628P on the risk of PD in ethnic Han-Chinese supported a significant association, and the odds ratio was 1.86. We further estimate the specific prevalence in relevant ethnic Han-Chinese subgroups. After stratifying the eligible data by gender, onset age, or family history, significant associations were found in all male, female, early-onset, late-onset, familial and sporadic subgroups, and the odds ratio were 1.90, 1.94, 2.12, 1.75, 6.71 and 1.81 respectively. In conclusion, our meta-analysis suggests that R1628P variant of LRRK2 has a significant association with the risk of PD in ethnic Han-Chinese and subgroup population. Parkinson’s disease (PD), characterized by the progressive and selective degeneration of substantia nigra dopaminergic neuron, is one of the most common neurodegenerative disorder in the population aged 65 years or older. Although PD was investigated intensely for ages, the pathogenesis of PD still remains indistinct. Genetic and environmental factors maybe play an interactional role in the etiology of PD. However, with the rapid growth of recent studies, genetic factors play a more and more important role in the progression of PD. Over the course of the past decades, some genes increase the risk of PD such as α-synuclein (SNCA), parkin PARK2), PTEN-induced putative kinase 1 (PINK1), oncogene DJ-1 (DJ-1), leucine-rich repeat kinase 2(LRRK2) and ATPase type 13A2 (ATP13A2) have emerged from previous investigations1, PTEN-inet al. It’s worth nothing that all confirmed pathogenic variants are various in diverse population and different region. The R1628P and G2385R of LRRK2 have been reported to be the most important single nucleotide polymorphism sites that increase the risk of PD in ethnic Han-Chinese populations7101315517Although hundreds of SNPs in LRRK2 are shown in PubMed database, only a few have been verified lead to autosomal dominant PD in the worldwide. The common variants of LRRK2, mainly distribute nearby the kinase domain, include A419VR1628P , within the COR domain, was found as the critical genetic risk factor for PD especially among ethnic Han-Chinese population in many previous studies71113Totally, 14 studies7111315202122232425262728−5) and heterogeneity test for whole studies indicated no significant between-study heterogeneity . Furthermore, the Begg’s test and Egger’s test were also implied no significant heterogeneity  with 9528 patients and 8707 controls examined the association between the R1628P variant and PD among ethnic Han-Chinese population. The main results  were perogeneity . Thus, s2 = 11.21, df = 4, P = 0.02, I2 = 64%) and OR = 1.94  < 10−4. Heterogeneity: χ2 = 5.14, df = 4, P = 0.27, I2 = 22%), respectively.We next analysis the available studies7112526We have got together 5 studies71125262 = 2.16, χ2 = 2.48, df = 1, P = 0.12 > 0.10, I2 = 60%) with 82 cases and 602 controls, 1.81  <10−5. Heterogeneity: χ2 = 14.41, df = 10, P = 0.15 > 0.10, I2 = 31%) with 7443 cases and 6661 controls.Only 2 studies77111520212324252627In the current study, we performed a comprehensive meta-analysis to clarify the epidemiological characteristics between R1628P variant and PD among ethnic Han Chinese population, and the specific prevalence in relevant ethnic Han Chinese subgroups, stratified by gender, onset age, or family history. A total of 14 eligible studies were included in the meta-analysis involving 9528 cases and 8707 controls. The literature selection process was showed in As our previously published study2-based Q-statistic, I2 index and Z score; random effect model was performed if I2 index is more than 50%. Only two studies of familial PD were included in this analysis, and the I2 index of between-study heterogeneity was more than 50%. The reliability of OR in familial PD subgroup was relatively low, and larger well-designed studies should be encouraged to elucidate the potential pathogenesis in familial PD with R1628P variant. Nevertheless, given over one billion population base of ethnic Han-Chinese, the carriers of R1628P variant are approximate 40–50 million, and precise elucidation of distribution patterns and molecular pathology of R1628P variant is still tremendously profitable to the personalized medical intervention of PD in ethnic Han-Chinese population.Two limitations should be mentioned in our meta-analysis results. Firstly, all the included studies focused on pathogenic LRRK2 R1628P variant were performed by clinical-based case-control studies. Although case-control studies are relatively simple to conduct and do not require a long follow-up period, the community-based cohort studies provide the best information about the causation of disease. Secondly, in the meta-analysis, between-study heterogeneity was tested by the χWe searched PubMed, Web of Science, Medline up to January 1th 2016 for all English language publications, CNKI  and CQVIP database for all Chinese language studies by using the following search terms: [(LRRK2 OR “leucine-rich repeat kinase 2” OR PARK8) AND (R1628P OR rs33949390 OR “c.4883G > C”) AND (“PD”)]. Studies using overlapping samples were excluded. In addition, case reports, reviews and editorials were not included.Only those studies assessing the association between LRRK2 R1628P variants and PD were included. For inclusion, eligible studies had to meet all of the following criteria: (1) be a case control or cohort study; (2) be published in a peer-reviewed journal; (3) have original data being independent from other studies, if the authors come from the same affiliation or an author has at least two publications, the corresponding authors were contacted to confirm whether the data was published repeatedly; (4) the subjects of control group has no history of mental disorders. For exclusion, the criteria as follows: (1) articles without case-control studies such as retrospective studies, reviews, and case reports; (2) other disease such as Alzheimer’s disease, Huntington’s disease; (3) studies have no sufficient data to calculate the odds ratio (OR) with its 95% confidence interval and p value.All the studies were screening by two independent reviewers. The disagreement of some studies was resolved through evaluating the records by the Newcastle-Ottawa Scale  and discussion. The following features were collected form all the including studies: first author, year of publication, ethnicity, location, genotyping method, numbers of case and control group and Hardy-Weinberg Equilibrium . And col2-based Q-statistic and quantified by I2 as a measure of the proportion of variance between the study-specific estimates that is attributable to between-study difference rather than random variation. Z score was conducted to test the overall effect and the significance set at P ≤ 0.05 (two-tailed test). The heterogeneity is considered statistically significant if PQ ≤ 0.10, we used the DerSimonian and Laird random effect model (D-L REM); otherwise, the Mantel-Haenszel fixed effect model (FEM) was used. All the forest plots and Begg’s funnel plot were display using Review Manager Software 5.2 (The Cochrane Collaboration). Sensitivity analysis was performed by leaving one out method to assess the stability of the results. Publication bias in each meta-analysis was detected by Begg’s funnel plots and Begg’s test (Stata v12.0).In all analysis, odds ratio (OR) with its 95% confidence interval (CI) was used to assess the strength of association between the LRRK2 R1628P variants and PD risk. Between-study heterogeneity was tested by the χIn addition, subgroup analysis was performed according to gender, onset age and family history for all studies. In the analysis of the onset age of PD for R1628P, we defined EOPD/LOPD (cutoff point: 50 years) according to the commonly used standard. 95% CIs were constructed using Woolf’s method. The Z test was used to determine the significance of the pooled OR.How to cite this article: Zhang, P. et al. Association of LRRK2 R1628P variant with Parkinson’s disease in Ethnic Han-Chinese and subgroup population. Sci. Rep.6, 35171; doi: 10.1038/srep35171 (2016).Publisher’s note: Springer Nature remains neutral with regard to jurisdictional claims in published maps and institutional affiliations.


Conversion to everolimus is often used in kidney transplantation to overcome calcineurin inhibitor (CNI) nephrotoxicity but there is conflicting evidence for this approach.To investigate the benefits and harm from randomized clinical trials (RCTs) involving the conversion from CNI to everolimus after kidney transplantation.Databases were searched up to March 2016. Two reviewers independently assessed trials for eligibility and quality, and extracted data. Results are expressed as risk ratio (RR) or mean difference (MD) with 95% confidence intervals (CI).2 greater than patients remaining on CNI (p = 0.0005) and the longer-term results (> 1 year) of renal function was identical to that of 1 year. There was not a substantial difference in graft loss, mortality, and the occurrence of adverse events (AEs) or serious AEs. However, the risks of acute rejection and trial termination due to AEs with everolimus are respectively 1.82 and 2.63 times greater than patients staying on CNI at 1 year posttransplant . Further, those patients who converted to everolimus had a substantially greater risk of anemia, hyperlipidemia, hypercholesterolemia, hypokalemia, proteinuria, stomatitis, mouth ulceration, and acne.Eleven RCTs, with a total of 1,633 patients, met the final inclusion criteria. Patients converted to everolimus had improved renal function at 1 year posttransplant with an estimated glomerular filtration rate (eGFR) of 5.36 mL/min per 1.73 mConversion from CNI to everolimus after kidney transplantation is associated with improved renal function in the first 5 years posttransplant but increases the risk of acute rejection at 1 year posttransplant and may not be well endured. Kidney transplantation is the treatment of choice for most patients with end-stage renal disease. Strategies to increase donor organ availability and to prolong the transplanted kidney’s survival have become priorities in kidney transplantation. Immunosuppressive therapy is essential and significant in this respect, nevertheless, but choosing the best suitable immunosuppressive therapy is still fairly complex. The calcineurin inhibitors (CNIs) are the principal components of immunosuppressive therapy after kidney transplantation and have made a major contribution to current long-term transplant outcomes, 2. MeanEverolimus (EVR), a mammalian target of the rapamycin inhibitor (mTORi), works 100 times greater than ciclosporin. It is part of a distinguished group of immunosuppressive drugs that have a divergent mode of action to that of CNIs even though they bind to the identical intracellular immunophilin as tacrolimus, namely, FKBP12. The mTORi/FKBP12 complex binds to and hinders the TORC1 complex, preventing proliferation of numerous cell types, such as T-lymphocytes, alloprimed B cells, and CD8 TAb-supp cells . The sidUp to now, there is no published work that systematically reviews EVR used as a maintenance therapy in renal transplant recipients. We undertook the meta-analysis of randomized controlled trials (RCTs) to systematically identify and summarize the current available evidence of the short- and long-term benefits and harm of conversion from CNI to EVR-based maintenance immunosuppression for kidney transplant recipients.A systematic literature search was performed using Pubmed/MEDLINE, OVID/EMBASE, and Cochrane Library covering the period from the inception of the database until March 2016 using a predefined algorithm  without Every RCT examining the conversion from CNI to EVR-based maintenance immunosuppression in adult isolated kidney transplantation was assessed. Research was judged to be suitable if they examined sudden or slow conversion to EVR, in the initial or later kidney transplant recipients, regardless of the time posttransplantation and baseline renal function. Trials that were deemed suitable included those in which the intervention (conversion to EVR) and reference (CNI continuation) groups were administered more maintenance immunosuppression comprising antimetabolites (mycophenolate or azathioprine) and steroids. Observational and uncontrolled studies, those examining children, those in which the participants obtained other solid organs plus a renal transplantation, and animal trials were eliminated.Data removal was done separately by 2 investigators (Jinyu Liu and Ruxu You) and discrepancies were resolved by a third investigator (Juan Li). We documented data on trial characteristics and demographics, such as authors’ names, year published, journal’s name, sample size per-arm, population characteristics, indications, CNI dose, EVR characteristics  and length of the trial. The main meta-analysis result was renal function  and secondary results were acute rejection, graft loss, patient survival and AEs. All of the results were investigated 1 year posttransplantation; long-term results, if available, were also investigated.The methodological quality of the included trials and the risk of bias were evaluated with elements from the Cochrane Collaboration’s tool for determining the risk of bias , 20. The2 (ranging from 0% to 100%). A Cochran’s Q p < 0.10 was considered to show significant heterogeneity, and analysis was undertaken using the random effects model. Otherwise, the fixed effects model was used. Summary effect estimates for renal function, acute rejection, graft loss, and patient survival were set on intention-to-treat (ITT) groups as identified in the study publications; further, all qualified trials documented AEs in the safety population and, therefore, summary effect sizes for AEs were approximated using safety populations as stated in the study publications.For every study we recovered, we computed the crude risk ratio (RR) estimates and corresponding 95% confidence intervals (CI) for the evaluated binary results , and the mean difference (MD) and corresponding standard deviation (SD) for the evaluated continuous outcomes (GFR). The existence of statistically significant heterogeneity was evaluated using the Cochran Q test and the degree of the observed heterogeneity was evaluated by the IPre-determined subgroup analyses were done to investigate origins of heterogeneity; trials were stratified by time of conversion to EVR  or by the duration to follow up. Sensitivity analyses were done based on to the protocol examining GFR or acute rejection. To additionally determine moderators of the noted effect estimates, a meta-regression analysis was performed to account for baseline renal function. Publication bias was investigated using funnel plots. Where information was missing we contacted the researchers to seek the pertinent information. Analyses were done in RevMan 5.3  and STATA 10 . All p values were 2-tailed. The study is documented based on the favored reporting items for systematic reviews and meta-analyses checklist .We identified 3,129 possibly relevant references . After examining the titles and abstracts and eliminating identical publications, 67 possibly qualified articles were determined. Fifteen reports –36 of 11All of the incorporated studies were designed to determine the safety and efficacy of conversion from CNI to EVR in adult kidney transplant patients. Design characteristics, immunosuppression regimens, and reported results for every study are summarized in The Cochrane Collaboration tool was implemented to evaluate the risk of bias . All tri2, 51%), , , and ; 2, 0%) compared to the early conversion trials , with reduction in heterogeneity . We performed additional sensitivity analysis excluding 2 trials [2, 42%). Further, meta-regression analyses taking into consideration the baseline GFR calculations revealed that the baseline renal function did not have a substantial impact on the difference in renal function between EVR and CNI groups 1 year posttransplantation.To additionally confront heterogeneity for renal function at 1 year after conversion to EVR, sensitivity analyses were done using the estimating procedures of GFR. We performed sensitivity analysis excluding 2 trials , 35 that2 trials , 24 that2, 43%; 2, 50% for patients converted to EVR). However, patients converted to EVR had a similar risk of acute rejection compared with the patients remaining on CNI up to 2, 3, and 5 years after transplantation [, , and ; 2 = 0%, p = 0.59).Every one of the included studies added to the meta-analysis examining the link between conversion from CNI to EVR and acute renal allograft rejection. Every one of the trials implemented the description of biopsy-proven acute rejection (BPAR), excluding the trial by Seckinger et al . In cont2, 37%; There were 3 trials , 28, 31 2, 0%; Mortality up to 1 year posttransplantation was reported in 8 studies. Altogether, 4 (0.62%) patients in the EVR group and 6 (0.95%) patients in the CNI group died within the first year after transplantation. There were no distinctions in mortality among patients who converted to EVR and those who continued on CNI .Five studies , 31, 33 2, 0%), and ; 2, 69%).Every one of the trials involved in the meta-analysis documented adverse events (AEs), although there were differences between studies in the nature and incidence of the reported adverse events. The main AEs involved in infection, blood and lymphatic system disorders, gastrointestinal disorders, metabolism and nutrition disorders, and the body as a whole-general disorders. Overall, 427 (94.05%) patients in the EVR group and 403 (90.77%) patients in the CNI group suffered 1 or more AEs within the first year after transplantation. There were no differences in the incidence of AEs or serious AEs between patients converted to EVR and those remaining on CNI [ were anemia, diarrhea, peripheral edema, proteinuria, mouth ulceration, pyrexia, urinary tract infection, upper respiratory tract infection, increased serum creatinine, cough, rash, and acne; the first 5 were substantially elevated in the EVR group than the CNI group (p < 0.05). In addition, leukocyte and platelet counts were decreased and liver enzymes  and lipids  were substantially elevated in the EVR group versus the CNI group (p < 0.05). No AEs were documented less commonly in the EVR group with a statistical difference.The findings from this meta-analysis establish that conversion from CNI to EVR-based maintenance immunosuppression after kidney transplantation is associated with an increased eGFR  up to 5 years after transplantation. There were no substantial differences in efficacy results between the 2 groups except that recipients who converted to EVR had a higher risk of acute graft rejection up to 1 year after transplantation. In terms of safety and tolerability, results were subordinate in the EVR group in contrast to the CNI group.2 (p = 0.0005) at 1 year and 6.50 mL/min per 1.73m2 (p = 0.002) 5 years after transplantation. Given the observed marked heterogeneity  for trials reporting on the effect of EVR conversion on renal function at 1 year after transplantation, caution is required with respect to the magnitude of the overall estimate for this outcome. Subgroup and sensitivity analyses reinforced the overall conclusion that conversion to EVR was associated with improved eGFR and eliminated heterogeneity as well. Nevertheless, those patients who converted to EVR had a substantially greater risk of proteinuria  at 1 year after the transplant and 2 patients developed proteinuria >1,000 mg/day post-conversion to everolimus. Offset against the insistent eGFR advantage of conversion to everolimus, this risk would seem to be allowable but because patients with elevated urine protein (e.g.> 150 or > 500 g/day) at the time of conversion seem to be more susceptible to gradual proteinuria after conversion to EVR [In renal transplantation, chronic allograft nephropathy (CAN) continues to be the primary reason for graft loss, 38. CANn to EVR , the traIt has been reported that conversion from CNI to mTORi increases the risk of acute rejection , 22. TheMammalian target of rapamycin inhibitors are associated with a number of well-described side effects , 13, 18 A longitudinal cohort study included 9,353 kidney transplant recipients showed that mTORi use was associated with a higher risk of all-cause mortality but not allograft loss . The majThere are further possible restrictions in the current meta-analysis. First, we did not incorporate any unpublished evidence, so there may be publication bias in the meta-analysis. Second, even though randomized evidence is safeguarded from selection bias, performance and detection biases could be possible confounders. An approach toward addressing this would be to exempt open-label trials. Ultimately, the majority of the incorporated trials were, by necessity, open-label trials, and so trial exclusion was not a viable choice. Lastly, a group of study parameters could possibly impede our study outcomes. The variation in baseline renal function between studies, and the unmistakable patient characteristics inside separate studies may have added to the noted heterogeneity.The currently available randomized evidence indicates that conversion from CNI to EVR after kidney transplantation is associated with short and long-term improvements in GFR in a number of studies, but these findings must be balanced against the greater rates of proteinuria, discontinuation due to AEs, and higher incidences of acute rejection up to 1 year posttransplant with the use of EVR as compared with CNI. Moreover, late conversion to EVR might be associated with a trend toward better renal function and a lower risk of acute rejection up to 1 year posttransplantion than early conversion. Additionally, in deciding the optimal immunosuppression strategy for their patients, clinicians should be alert to the increased risk of bone marrow suppression, dyslipidemia, mouth ulceration, and stomatitis after conversion to EVR which, however, has no effect on graft and patient survival.S1 Fig(TIF)Click here for additional data file.S2 Fig(TIF)Click here for additional data file.S1 Table(DOCX)Click here for additional data file.S2 Table(DOCX)Click here for additional data file.S3 Table(DOC)Click here for additional data file.S1 Text(DOCX)Click here for additional data file.S2 Text(DOC)Click here for additional data file.


Objective: To perform a dose–response meta-analysis of prospective cohort studies investigating the association between long-term coffee intake and risk of hypertension. Methods: An online systematic search of studies published up to November 2016 was performed. Linear and non-linear dose–response meta-analyses were conducted; potential evidence of heterogeneity, publication bias, and confounding effect of selected variables were investigated through sensitivity and meta-regression analyses. Results: Seven cohorts including 205,349 individuals and 44,120 cases of hypertension were included. In the non-linear analysis, there was a 9% significant decreased risk of hypertension per seven cups of coffee a day, while, in the linear dose–response association, there was a 1% decreased risk of hypertension for each additional cup of coffee per day. Among subgroups, there were significant inverse associations for females, caffeinated coffee, and studies conducted in the US with longer follow-up. Analysis of potential confounders revealed that smoking-related variables weakened the strength of association between coffee consumption and risk of hypertension. Conclusions: Increased coffee consumption is associated with a modest decrease in risk of hypertension in prospective cohort studies. Smoking status is a potential effect modifier on the association between coffee consumption and risk of hypertension. Coffee has been the focus of major attention due to its global consumption and impact on health . HistoriOverall, the benefits of coffee on human health seem evident; however, the association between coffee consumption and in particular risk of hypertension remains open for debate. Results from meta-analyses of randomized controlled trials testing the effects of caffeine intake and decaffeinated coffee on blood pressure were null and showed no increase in blood pressure ,12; howeThe design, analysis, and reporting of this study is compliant with the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) guidelines .http://www.ncbi.nlm.nih.gov/pubmed/) and EMBASE (http://www.embase.com/) databases was conducted for studies up to November 2016 using the following search strategy: (“coffee” (MeSH Terms) OR “coffee” (All Fields)) AND (“hypertension” (MeSH Terms) OR “hypertension” (All Fields)) AND (cohort (All Fields) OR  OR  AND “studies” (All Fields)) OR “longitudinal studies” (All Fields) OR “prospective” (All Fields) OR cases (All Fields). The inclusion criteria consisted of studies that: (i) are prospective; (ii) evaluated the association of coffee intake and the risk of hypertension in individuals without hypertension at baseline; (iii) assessed and reported hazard ratios (HRs), risk ratios (RRs), or odds ratios (ORs) and 95% CI for hypertension; and (iv) assessed and reported a defined amount of coffee consumption  for each category of exposure. Reference lists of the included studies were also searched for any additional study not previously identified. For papers that studied the same cohort, only the study that included the entire cohort or had the longest follow-up was included.A systematic search of PubMed (Data were extracted by two independent investigators (G.G. and A.M.) using a standardized extraction form. Discrepancies on the included studies were discussed and resolved by consensus. The following information was collected: (i) first author name; (ii) year of publication; (iii) study cohort name and country; (iv) total number, sex, and age (mean or range) of participants; (v) follow-up period; (vi) distributions of cases and person-years, HRs/RRs/ORs and 95% CIs for all categories of exposure; (vii) coffee intake range for each category of exposure; (viii) adjusted covariates; (ix) prevalence of smokers, diabetic subjects, and participants with low physical activity for each category of exposure; and (x) baseline mean/median systolic and diastolic blood pressure, age, body mass index (BMI), and daily intake of sodium and potassium for each category of coffee intake.The quality of each study was assessed according to the Newcastle-Ottawa Quality Assessment Scale  composedIn this meta-analysis, ORs and HRs referring to new onset incident cases of hypertension were deemed equivalent to risk ratios (RRs) . When cop-value for non-linearity by testing whether the coefficient of the second spline was equal to zero. A sensitivity analysis by excluding one study at the time was conducted to determine the stability and robustness of the results.First, generalized least-squares (GLS) were used to calculate study-specific coefficients across categories of coffee intake accounting for the correlation within each set of RRs ,18. Non-To test for potential confounders, subgroup analyses by sex, type of coffee, and geographical area were performed. To further investigate whether other unmeasured potential confounders should be included in the interpretation of the results, we investigated the distribution of systolic blood pressure, diastolic blood pressure, age, BMI, daily intake of sodium and potassium, percentage of smokers, participants with low physical activity and those with type-2 diabetes across categories of coffee consumption. For this purpose, a separate two-stage bivariate meta-analysis was performed for each confounder variable to determine its association with coffee intake . First, Q-test tested for statistical heterogeneity  and quantified through the multivariate generalization of the I2 statistic . All analyses were performed on R, software version 3.0.3, dosresmeta and mvmeta packages .Publication bias was assessed using Egger’s regression test while the Cochran The selection process of studies included in the meta-analysis is illustrated in I2 = 51%, heterogeneityP = 0.004) and no publication bias (EggerP = 0.755)  . After e= 0.755) . When li= 0.755) B, where The distribution of systolic and diastolic blood pressure and percentage of smokers across categories of coffee consumption was explored in four studies where a significant association was evident only with diastolic blood pressure . The preThis meta-analysis illustrated a linear association between increasing intake of coffee consumption and decreased risk of hypertension in six prospective cohort studies. However, when smoking status was included as a moderator in the meta-regression analyses, the association was weakened, suggesting that smoking may be an effect modifier. Compared to previous meta-analyses ,14, thisFrom the subgroup analyses, this study showed significant results in studies (or datasets) that only included women, were conducted in the US, and had a longer follow-up. Significant results seen in women and not in men could be attributed to increased prevalence of smoking in males accounting for 30% of all mortality in males 50–70 years . To accoApart from those above, other reasons may account for the differences observed between subgroups , including variations in preferred coffee constituents, variations in brewing and in the actual amount of coffee contained in “a cup” across regions, and differences in population genetics. There is evidence that genetic factors related to target receptors and CYP1A2, such as the genetic polymorphism rs762551, also known as −136 C>A CYP1A2, encoding the CYP1A2*1F allele of the CYP1A2 gene, leads to slow metabolism of caffeine ,30. CarrAnother finding of this study was the potential effect modification of sodium intake on risk of hypertension associated with coffee consumption. Despite insufficient data for further examining this issue, three studies showed direct association between sodium intake and coffee consumption where increased intake of one cup of coffee per day was associated with 38 mg/day additional intake of sodium. It is known that increased sodium intake may be associated with increased fluid retention, but does not increase urine volume excretion, thus leading to a rise in body weight and blood pressure . A comprThe biological evidence supporting the association between coffee consumption and decreased risk of hypertension discounting caffeine intake has been investigated only during the last decade. The coffee content in phenolic compounds has been considered the main factor responsible for the beneficial effects on blood pressure mentioned previously . The mosThis meta-analysis has some limitations that should be addressed. The main issue in interpreting our findings depends on the potential confounders  and genetic polymorphisms that are not explored in existing cohorts. We attempted an in-depth analysis on the role of smoking status, suggesting that it poses as an effect modifier. No data on genetic polymorphisms and only limited data on sodium intake were available and, hence, we were not able to analyze the potential effect modification of these variables, which may have weakened the observed results. Nevertheless, the large number of individuals included, the lack of evidence on heterogeneity, and the observed significant inverse association observed for the main effect variables all support the robustness of the findings. Another limitation in the papers included in this study is the lack of information on the type of coffee used  and on the actual amount of coffee used in one cup. The type of coffee consumed by US cohorts could be different from those consumed by European cohorts, which contribute to the heterogeneous findings. Similarly, the amount of coffee added in a cup can vary across countries. One cup of coffee could refer to 150 mL or 300 mL which is another source of heterogeneity in these results. However, a previous meta-analysis that adjusted for the varying measurements of coffee showed identical results to the unadjusted measures on the outcome investigated .Considering that the results from this meta-analysis may not account for the genetic variations contributing to the slow metabolism of caffeine, overall current evidence suggest that usual coffee consumption in the long-term is not a risk factor for developing hypertension in existing cohort studies. Rather results show an inverse association of increase coffee consumption and the risk of developing hypertension. Due to lack of long-term randomized controlled trials, results from cohort studies represent the most reliable available evidence when studying long-term effects of food on future health.To date, our study provides a robust overview of the highest level of evidence that currently exists on the association of coffee consumption and risk of hypertension. Nevertheless, more cohort studies are needed to provide more in-depth and long-term evidence of this association. Additional factors such as smoking status, sodium intake and genetic polymorphisms need to be addressed in future studies to allow for stratified sub-group analyses. Further evidence on decaffeinated coffee could also provide information on the role of caffeine in such long-term outcomes.


Much of the related researches have reported the correlation between renal transplantation and different tumors in the post transplant recipients. However, there are not exact essays revealed that renal transplantation is definite causation for liver carcinoma, thus we systematically evaluated the association between renal transplantation and the risk of liver carcinoma in this meta-analysis from all available researches.All useful data were collected through searching of PubMed and Web of Science until the date of 31 September 2015. Random-effects model were adopted to calculate the standardized incidence ratio and 95% confidence interval (CIs) of the risk of liver carcinoma among renal transplant recipients. Other statistical analyses like heterogeneity tests, sensitivity analysis and publication bias were also performed in this meta-analysis.P=0.005) of liver carcinoma following renal transplantation compared with the general population. Observation and publication bias were not observed in this study.Among 17,4256 kidney transplant cases and 25,6736 patients-years observation, 9136 post-transplant cancers were diagnosed. We identified a 2.08-fold higher standardized incidence rate (SIR) : 1.24-3.47, This study suggested that the risk of liver carcinoma among renal transplant recipients with chronic hepatic disease is higher than general population. Such results alert clinical doctors the importance of anti-virus therapy with chronic virus hepatitis and enough attention of periodic liver screening with chronic liver diseases in renal transplant recipients. Chronic kidney disease (CKD) is quite a common disease around the world, its etiology from infectious disease to emerging diseases especially hypertension and diabetes –3. CKD ide novo malignancy is the only sideline of this population, since tumor-specific mortality is not the main reason of the population [Renal transplantation is regarded as the therapeutic option of choice in the end-stage renal failure . Initialpulation . Howeverpulation , 9. At tpulation , 11.Hepatocellular carcinoma (HCC) is also a common tumor worldwide, it’s popularity has related geographic differences. Epidemiology revealed that areas with high incidence include Africa, in Asia especially southeastern countries , however, in majority areas of China and Japan. Its main risk factors are hepatitis virus infection like hepatitis B and hepatitis C, the other susceptible factors are alcohol consumption, aflatoxin B1 intake and other associated congenital diseases .Some studies have showed the increased risk of liver carcinoma after renal transplantation , 14. HowA total of 9 studies met our inclusion criteria , 15–22. Table 2I=89.2%, P heterogeneity=0.000) in the pooled analysis.This meta-analysis for the SIR for liver cancer suggested a significantly increased risk compared to the common population  and 2.428. Tumors such as NHL and Kaposi’s carcinoma, which are rare in the general population, however, are encountered more frequently in solid organ transplantation recipients . Liver cThe exact mechanisms of the oncogenesis in renal transplantation recipients are still unknown. Several possible related factors had been proposed. One hypothesis is that cancers result from related viral infection(hepatitis B virus or hepatitis C virus related) and another is reduced immune surveillance against a variety of non-viral tumor antigens . For a pThere are some limitations in this meta-analysis. 1) Despite the funnel plot and Egger’s test showed the absence of publication bias, some important studies except for English publications or unpublished date related to our studies may have been missed, especially when the pooled increased estimate for liver cancer after transplantation was generated from the limited publications. 2) This meta-analysis included several different genetic background, medical situation, life styles, no detailed information for us to perform an adjustment for these potential confounders, such as the period of chronic hepatitis virus carrier, the period of alcoholic liver cirrhosis, obesity and so on. 3) In our collected data, although all studies used the general population as the reference population, the criteria used for matching might be different applied, 4) In our screened studies, two important indices, mean age at diagnosis of malignancy and mean time to develop liver cancer, were not measured in the included studies. 5) The renal transplant recipients were not screened for liver cancer before kidney transplantation . What’s more, the length of time from transplantation to the diagnosis of the liver cancer was not clear in our results. So we could not exclude the recipients of possible pre-existing tumors in these recipients. 6) Related information on the type and the dosage of immunosuppressive agents were not available for the further subgroup analysis. Some studies demonstrated that low-dose cyclosporine was associated with a lower rate of malignancy than high-dose cyclosporine in RTRs .In designing our meta-analysis, a systematic and comprehensive PubMed literature search and Web of Science Databases search updated through September 2015 was conducted to identify studies involving the risk of liver cancer in renal transplantation recipients. The search was restricted to studies published in English and to those including only humans. Search key terms used were ‘renal transplantation’, ‘kidney transplantation’, ‘liver’ and ‘cancer’. The search results were restricted to the presence of the key terms in the title, abstracts and unpublished reports were excluded.Eligible studies were selected for meta-analysis if they conform to the following inclusion criteria:(1) studies must be population-based cohort studies in renal transplantation recipients; (2) SIR were calculated with 95% confidence intervals in transplant recipients as compared to the general population. (3) Sufficient data on the incidence rate, SIR or relative risk (RR) of liver cancer and enough numbers of patients should be included in the studies. The following exclusion criteria were used: (1) other organ transplantation studies; (2) they were case series, case-control studies, or case reports; (3) they evaluated liver cancer following transplantation of organs other than the kidney; or (4) they evaluated cancers other than liver cancer following renal transplantation.The final date from all qualified publication were independently by two investigators (PC and TW) using the inclusion criteria above. They reached an agreement on all items after reducing any form of bias. Important information extracted from each study included: the first author, publication year, the type of transplantation, data source, geographic origin, number of patients, number of renal transplantation cases, length of follow-up time, number of all cancers in kidney transplant cases, mean follow-up time(years), patients-years(years), mean age at transplantation(years), number of expected cases of liver cancer, number of identified cases of liver cancer, and the SIRs of commonly known cancers and liver cancer.2I, which was documented for the percentage of the observed between study variability due to heterogeneity rather than chance, with 2I<25%, 25%-75% and >75% representing low, moderate and high degree of inconsistency, respectively.In this meta-analysis, we obtained an estimate from each study of the unadjusted relative risk (RR) with 95% confidence intervals (CIs) to assess the strength of liver cancer in renal transplantation recipients compared to the general population. Because of possible heterogeneity in our studies, a random-effects model was used to pool effects for RR . HeterogWe assessed the effects of individual study data on the pooled RR through one-way sensitivity analysis. After the sequential removal of each study, the remaining studies could be used to evaluate the stability of the results. At the same time, we assessed publication bias by using a funnel plot and Begg’s test to find out whether there was a bias towards publication of studies . Funnel In conclusion, our meta-analysis showed an increased risk of developing liver carcinoma in transplant recipients . Renal transplant recipients should therefore be screened cautiously for their liver function and adjusted the dosage or period of immunosuppresive agents according to patients’ clinical examination results. If they are the recipients with chronic hepatitis, periodic ultrasound and alpha fetoprotein monitering every 6-12 months are recommended among chronic hepatitis carriers because early detection of tumor can have a higher chance of receiving treatment .


Abdominal aortic aneurysm (AAA) is a life-threatening condition. A number of studies reported the association between methylenetetrahydrofolate reductase (MTHFR) C677T polymorphism and AAA risk, but substantial controversial findings were observed and the strength of the association remains unclear.The aim of this study was to investigate the aforementioned association in the overall population and different subgroups.PUBMED and EMBASE databases were searched until March 2016 to identify eligible studies, restricted to humans and articles published in English. Summary odds ratios (ORs) and 95% confidence intervals (CIs) were used to evaluate the susceptibility to AAA. Subgroup meta-analyses were conducted on features of the population, such as ethnicity, sex of the participants, and study design (source of control).P = 0.02), cases with prevalence of smoking <60% , and cases with aneurysm diameter ≥55 mm  in the dominant genetic model. No publication bias was detected in the present study.Twelve case–control studies on MTHFR C677T polymorphism and AAA risk, including 3555 cases and 6568 case-free controls were identified. The results revealed no significant association between the MTHFR C677T polymorphism and AAA risk in the overall population and within Caucasian or Asian subpopulations in all 5 genetic models. Further subgroup meta-analysis indicated that significantly increased risks were observed among cases with a mean age <70 years (OR = 1.73, 95% CI = 1.10–2.12, In conclusion, our comprehensive meta-analysis suggests that the MTHFR C677T polymorphism may play an important role in AAA susceptibility, especially in younger, non-smoking, larger AAA-diameter subgroups of patients Protective factors include type-2 diabetes mellitus, African-American ethnicity, and femaleness.,5,6 Although many risk factors have been identified, the etiology of AAA is still largely unclear.,8Abdominal aortic aneurysm (AAA) is a life-threatening condition detected in up to 9% of men ages >65 years in Western populations.in vitro, and has been associated with differences in Hcy concentration.,10 Previous studies on the MTHFR C677T polymorphism suggested a causal link between hyperhomocysteinemia (HHcy), characterized by elevated Hcy levels in the blood, coronary heart disease, stroke, and peripheral vascular disease.,12Homocysteine (Hcy) is a sulfur-containing nonessential amino acid that functions as a key intermediate in the methionine metabolism. Methylenetetrahydrofolate reductase (MTHFR) is a key enzyme involved in Hcy metabolism. A common functional polymorphism (C677T) in the gene-encoding MTHFR results in a 70% reduction in enzymatic activity, –20 Whether Hcy levels play a causal role in the pathogenic process of AAA remains unclear. To overcome the problem of reverse causality, studies have focused on the association between the MTHFR C677T polymorphism and AAA risk. However, the association is still inconsistent. Although 5 case–control studies –15,21,22 showed that carriers of the MTHFR 677T allele had an increased risk of AAA, 7 others,20,23–26 showed no significant association between these 2 factors. Furthermore, 4 studies,27–29 investigating this association using forest plots also provided conflicting results. The present study was designed to investigate the association between the MTHFR C677T polymorphism and AAA risk by performing a comprehensive meta-analysis, including a subgroup analysis.However, studies investigating the association between HHcy and AAA yielded conflicting results.22.1Two independent investigators performed a comprehensive search for all studies on the association between the MTHFR C677T polymorphism and AAA risk in 2 electronic databases (PubMed and Embase) through March 31, 2016. We used the search terms “mthfr,” “methylenetetrahydrofolate reductase,” “5,10 methylenetetrahydrofolate reductase,” “677C,” “C677T,” and “rs1801133” in combination with “aneurysm.” We searched for any additional studies in the references of all identified publications, including previous relevant meta-analyses. Unpublished data and supplemental information were also requested from authors. The search was restricted to studies in humans and articles published in English. This study protocol was approved by the ethics committee of Chinese PLA General Hospital.2.2Two investigators independently screened all titles and abstracts of the identified studies. Only studies published as full-length articles or letters in peer-reviewed journals were included. Studies included were in accordance with the following criteria: original case–control studies with matching control subjects; clinical trials studying the association between the MTHFR C677T polymorphism and AAA risk; all cases were diagnosed by ultrasound or computed tomography angiography; contained at least 2 comparison groups (AAA group and control group); all genotype distributions were reported in both case and control groups. Accordingly, the following exclusion criteria were also used: no controls; insufficient data reported ; and studies with duplicated data.2.3For each study, data were extracted using a standard form by 2 independent authors. For each included study, the following information was collected: name of the first author; year of publication; country of origin, ethnicity, and sex of the participants; study design (source of control); numbers of cases and controls; mean age in case groups; prevalence of smoking in case groups; mean diameter of aneurysms; and numbers of CC, CT, and TT genotypes in cases and controls. In the event of a disagreement between authors, discrepancies of included studies were resolved by a third author.2.4Z test, and a P value <0.05 was considered significant. We performed subgroup-analysis according to ethnicity (Caucasian or Asian), sex of the participants , source of control , mean age in case groups (≥70 years or <70 years), prevalence of smoking in case groups (≥60% or <60%), and mean diameter of aneurysm (≥55 mm or <55 mm).We carried out this meta-analysis referring to PRISMA guidelines. The associations between the MTHFR C677T polymorphism and AAA were calculated as the pooled odds ratio (OR) with its corresponding 95% confidence interval (CI) in the allele contrast (T vs C), homozygote (TT vs CC), heterozygote (CT vs CC), recessive (TT vs CT + CC), and dominant (TT + CT vs CC) genetic models. The significance of the pooled OR was determined by the Q test and the I2 statistic. Heterogeneity was regarded as statistically significant if Phet < 0.10 or I2 > 50%. The random-effects model was used if the test of heterogeneity was significant. Otherwise, the fixed-effects model would be applied in the analysis. To assess the stability of the results, sensitivity analyses were performed to determine whether the exclusion of any studies could affect the results. The potential publication bias was primarily appraised by Begg funnel plot and further evaluated using the Egger test  with STATA 12.1 software .Heterogeneity was assessed by the chi-square–based Cochran P values in the meta-analysis were 2-tailed, and P values <0.05 were considered significant.EpiData 3.1  was used to gather initial data in a standard form, and Endnote X7  was used for literature management. All statistical analyses were carried out with the review manager  and the STATA software version 12.1 . All 33.1Using the Pubmed and Embase databases, 119 studies were identified. After excluding 48 duplicates, 71 studies were identified for further evaluation. After title and abstract screening, 53 irrelevant studies were excluded. After assessing the remaining 18 full-text articles, 4 studies were excluded for insufficient data, and 2 for not being case–control studies . Ten studies,18,20–26 included both males and females and the remaining 2,17 included only males. In 6 studies,,20–22,25 the mean age of the case group was <70 years. In 5 studies,–15,21,26 the prevalence of smoking in the case group was <60%. In 6 studies,,21–23,25 the mean diameter of aneurysm was >55 mm.Ultimately, a total of 12 case–control studies,3.2P < 0.01).The frequency of the T allele varied widely across the 12 studies, ranging from 0.21 to 0.59 . Ten studies,18,21–26, including 2937 cases and 5795 controls, were used to investigate the association in Caucasian populations. The results showed no association between the MTHFR C677T polymorphism and AAA risk in Caucasians using all 5 genetic models . Five studies–15,21,26 with prevalence of smoking in the case group <60, including 1163 cases and 1204 controls, showed a significant association in the dominant model . Six studies,21–23,25 with AAA diameter ≥55 mm, including 1768 cases and 1695 controls, showed a significant association based on the dominant models .The subgroup meta-analyses are shown in Table 3.3P > 0.05).A Begg funnel plot and an Egger test were performed to assess the publication bias. As shown in Fig. 4,23 However, the results of previous studies,20,24,31 that investigated the relationship between HHcy and AAA are inconsistent. Whether HHcy plays a direct causal role in the pathogenesis of AAA remains elusive. To overcome the problem of reverse causality, several studies,18,20–26 sought an association between MTHFR genotype and AAA risk. However, the association between the MTHFR C677T polymorphism and AAA is still controversial.HHcy is an independent risk factor for the development of atherosclerosis, cardiovascular disease, and stroke.In the present meta-analysis with 3555 cases and 6568 controls from 12 independent studies, no significant association was detected between the MTHFR C677T polymorphism and AAA risk in the overall population in all 5 genetic models. Additionally, no significant association was revealed within Caucasian or Asian subpopulations. However, the subgroup meta-analysis showed a significant association between the MTHFR C677T polymorphism and AAA risk in the subgroups of cases with a mean age <70 years, prevalence of smoking <60%, and AAA diameter ≥55 mm in the dominant genetic model. The results suggest that age, smoking, and AAA diameter may play a role in the relationship between the MTHFR C677T polymorphism and AAA risk.,27–29 using forest plots investigated the association between the MTHFR C677T polymorphism and AAA risk, the results of these studies were controversial and lacked a subgroup analysis. Three meta-analyses,27,28 showed that carriers of the MTHFR 677T allele presented with an increased risk of AAA. The study by Thompson et al, including 1075 cases and 1067 controls from 5 independent studies, revealed a significant effect of the T allele variant in increasing the risk of AAA . McColgan et al, including 817 cases and 629 controls from 6 studies , showed a similar significant result . Saratzis et al, including 2498 cases and 6212 controls from 9 studies, also showed a significant OR between the MTHFR C677T polymorphism and AAA . The meta-analysis  by Bradley et al, including 2498 cases and 6212 controls from 9 studies, showed no significant association between the MTHFR C677T polymorphism and AAA , OR = 1.06; 95% CI = 0.81–1.40). Our study, including 3555 cases and 6568 controls from 12 studies also showed no significant association between the variant genotypes and AAA risk in the 5 different genetic models. Furthermore, our meta-analysis of 10 studies,18,21–26 in Caucasians, including 2937 cases and 5795 controls, showed no association between AAA risk and the variant genotypes in the different genetic models. Two studies in Asians,20 also showed no association in all genetic models. Our results indicated that there is no significant association between the MTHFR C677T polymorphism and AAA in the overall population (both Caucasians and Asians).Although 4 studiesHowever, the characteristics of included populations vary among different studies; the possibility of high heterogeneity and publication bias should be considered in a meta-analysis. Furthermore, a comprehensive meta-analysis requires the analysis of subgroups.,20–22,25 with a mean age <70 years in case groups, showed a significant association between the MTHFR C677T polymorphism and AAA risk in the dominant model . Ferrara et al divided the patients of the AAA group into 2 groups according to age and determined that the frequency of the 677T allele in younger patients with AAA was significantly higher than that in older patients with AAA. Our subgroup meta-analysis of 5 studies–15,21,26 of cases with prevalence of smoking <60% also showed a significant association between the MTHFR C677T polymorphism and AAA risk . Strass et al divided the patients of the AAA group into 2 groups—those who smoke and those who do not—and showed that the frequency of the 677T allele in patients with AAA who smoke was higher than that in patients with AAA who do not, but this was not significant. In our subgroup meta-analysis of 6 studies,21–23,25 of AAA with a diameter ≥55 mm, a significant association between the MTHFR C677T polymorphism and AAA risk was observed . Jones et al showed AAA diameter was significantly greater in T homozygotes compared to that in C allele carriers and the OR for T homozygotes and aneurysm size was 1.35 . However, 4 other studies,17,21,23 indicated that the aneurysm diameter was not significantly larger in the AAA patients with the MTHFR 677TT polymorphism than in those with the MTHFR 677CC polymorphism. Our previous study also showed no significant association between the MTHFR C677T allele and aneurysm diameter with no significantly higher increase in aneurysm size in patients with the CT or TT genotype than in those with the CC genotype. Interestingly, in our previous study, there was a significant association between the MTHFR 677T allele and the risk of AAA  in the subgroup of patients who consumed alcohol, indicating that consuming alcohol may play an important role in the association between the MTHFR C677T polymorphism and AAA. These results indicate that other factors may play a role in the association between the MTHFR C677T polymorphism and AAA. Additional studies focusing on the association between the MTHFR C677T polymorphism and AAA taking into account confounding factors and their interactions as well as subgroup analyses are needed to clarify these findings.In our subgroup meta-analysis, the subgroup of 6 studiesOur study provides evidence of the association between the MTHFR C677T polymorphism and AAA risk by performing a comprehensive meta-analysis. When interpreting our findings, several limitations should be considered. First, only 2 of the included studies analyzed Asian populations and no study data from ethnicities other than Caucasian and Asian populations were included. Thus, additional studies with large sample sizes designed to include other ethnicities are needed. Secondly, detailed information on individual patients in some studies was not available. Thus, we could not assess the risk of AAA according to stratification of other AAA risk factors. Gene–gene and gene–environment interactions were not fully resolved in this meta-analysis due to the lack of available data and could be addressed in future studies. Finally, our main analysis was based on unadjusted estimates of ORs owing to the lack of sufficient data. However, a more precise analysis should be conducted if adjusted estimates of ORs are available in all studies.5In summary, our present meta-analysis provides evidence of the association between the MTHFR C677T polymorphism and AAA risk, and suggests that the MTHFR T allele increases the risk of AAA in younger, nonsmoking, larger-AAA-diameter subgroups of patients. Future studies, including different subgroup analyses, are warranted to verify our findings.


VEGF) polymorphisms  and AMD risk, also stratifying for AMD subtypes and ethnicity. A systematic literature search in the Medline database, using PubMed, was carried out for epidemiological studies, published before June 2016. Associations of VEGF polymorphisms with AMD were estimated by calculating pooled odds ratios (ORs) and 95% confidence intervals (95% CIs) based on different models. Twelve articles were included in the analysis. The present meta-analysis constitutes a useful guide for readers to study AMD and adds new evidence to the growing literature on the role of VEGF polymorphisms in the risk of AMD. Significant associations with AMD risk were showed for rs833061, rs1413711, and rs3025039 polymorphisms but not for rs2010963.Age-related macular degeneration (AMD) is the most common cause of blindness in elderly people worldwide and the major degenerative disease of the retina that leads to progressive impairment of central vision. Several polymorphisms in different genes have been proposed as factors that increase the disease susceptibility. The aim of the present study is to carry out a systematic review and an updated meta-analysis in order to summarize the current published studies and to evaluate the associations between four common vascular endothelial growth factor ( Age-related macular degeneration (AMD), the most common cause of blindness in elderly people worldwide , is the ARMD1), apolipoprotein E (APOE), and complement factor H (CFH); vascular endothelial growth factor (VEGF) genes have also been reviewed [ VEGF gene is located on chromosome 6p21.3 and contains seven introns and eight exons [Epidemiological studies suggest that the pathogenesis of this complex disorder implicates sociodemographic (age and race), environmental , and genetic risk factors , 5. Partreviewed . VEGF hareviewed . The VEGreviewed , inducinreviewed , 14. Theht exons –17. VEGF gene and AMD risk. Among these SNPs, rs833061 (−460T/C) in the promoter region, rs1413711 (+674C/T) in intron 1, rs3025039 (+936C/T) in the 3′-untranslated region, and rs2010963 (+405G/C) in the 5′-untranslated region were found to be associated with the AMD susceptibility [ VEGF gene may have a critical role in determining AMD risk, evidence by studies on small or moderate sample sizes remain ambiguous. These contentious results were also reported by recently published meta-analyses [Recently, several studies have focused on the association between single-nucleotide polymorphisms (SNPs) in thetibility –20. Althanalyses –23. VEGF gene polymorphisms  and AMD risk, also stratifying for AMD subtypes and ethnicity.The aim of the present study is to carry out a more comprehensive systematic review and an updated meta-analysis in order to summarize the current published studies and to evaluate the associations between four common VEGF polymorphisms and AMD risk. The literature search, limited to studies written in English, was independently conducted by the two authors, using the following key words:  AND (“variant” or “SNPs” or “polymorphism”) AND (“age-related macular degeneration” or “AMD”). Moreover, the references from retrieved articles were also checked to search for additional studies.A systematic review of original articles, published before June 2016, was conducted by searching in the Medline database in order to collect data from epidemiological studies investigating the association betweenThe selection criteria were as follows: (1) studies must employ a case-control or cohort design (2) and must evaluate the associations between VEGF polymorphisms  and AMD; (3) genotype data of patients and controls must be available in order to estimate odds ratios (ORs) and 95% confidence intervals (95% CIs). Furthermore, exclusion criteria were (1) studies that did not provide genotype data in AMD patients and/or in control subjects and (2) review articles. The meta-analysis was conducted according to the preferred reporting items for systematic reviews and meta-analysis (PRISMA) guidelines .p value for Hardy-Weinberg equilibrium test in subjects without AMD.The following information were extracted: first author's last name, year of publication, country and ethnicity, sample size of subjects with and without AMD, genotype distribution of case and control groups, subtypes of AMD cases and controls (dry AMD and wet AMD), and http://ims.cochrane.org/revman), was used to estimate the association between VEGF polymorphisms and AMD risk.The REVIEW MANAGER 5.2 software, provided by the Cochrane Collaboration (p < 0.05 was considered significant).For each polymorphism, the wild-type allele was set as 1 and the risk allele as 2 and the Chi-square test was performed to determine if the genotype distribution of the control subjects is deviated from the Hardy-Weinberg equilibrium , the following genetic models were adopted for each polymorphism: the homozygote model (22 versus 11), the dominant model (22 + 12 versus 11), the recessive model (22 versus 12 + 11), the heterozygote model (12 versus 11), and the allelic model (1 versus 2) .Z test, and p < 0.05 was considered significant. Heterogeneity across studies was assessed using the Q-test based on the χ2 statistic .The significance of pooled OR was determined using the I2 value was calculated and interpreted as follows: an I2 value of 0% indicates “no heterogeneity,” whereas 25% is “low,” 50% is “moderate,” and 75% is “high” heterogeneity [τ2) statistic [To quantify heterogeneity, the ogeneity . The bettatistic .According to heterogeneity across studies, the fixed-effects (Mantel-Haenszel method) or random effects models (Der Simonian-Laird method) were used to calculate pooled effect estimates.Furthermore, subgroup analyses by subtypes of AMD and ethnicity (Asian and Caucasian) were conducted. The leave-one-out sensitivity analysis, by the omission of a single study at a time, was performed in order to assess whether a particular omission could affect the overall OR value and the heterogeneity across studies. To identify potential publication bias, the asymmetry of the funnel plots, in which ORs were plotted against their corresponding standard errors, was examined.The detailed steps of the systematic review and meta-analysis process are given as a PRISMA flowchart in For the rs833061 polymorphism, the meta-analysis showed a significant association with AMD under a homozygote model ; a dominant model ; and an allelic model . Pooled ORs, under a heterozygote model  and a recessive model , were not statistically significant. Subgroup analysis by ethnicity confirmed that, in Asians, the polymorphism was associated with AMD under the homozygote , the recessive , and the allelic  models; in Caucasians, the polymorphism was associated with AMD under the homozygote  and the allelic  models.Subgroup analysis by subtypes of AMD confirmed that this polymorphism was associated with wet AMD under a homozygote model  ; a dominFor the rs1413711 polymorphism, the meta-analysis showed no significant association between the polymorphism and AMD under a homozygote model ; a dominant model ; a heterozygote model ; a recessive model ; and an allelic model . However, the stratified analysis indicated that this polymorphism was associated with wet AMD under a recessive model  . SubgrouFor the rs3025039 polymorphism, the meta-analysis showed that no significant association between the polymorphism and AMD was assessed under a homozygote model ; a dominant model ; a heterozygote model ; a recessive model ; and an allelic model . Further subgroup analyses, by subtypes of AMD and ethnicity, confirmed that there was no association between this polymorphism and the risk of AMD in any of the genetic models.For the rs2010963 polymorphism, the meta-analysis showed that no significant association between the polymorphism and AMD was assessed under a homozygote model ; a recessive model ; a heterozygote model ; a dominant model ; and an allelic model . Further subgroup analyses, by subtypes of AMD and ethnicity, confirmed that there was no association between this polymorphism and the risk of AMD in any of the genetic models.The leave-one-out sensitivity analysis was performed in order to investigate the sources of heterogeneity across studies.Q-test and I2 statistics showed no significant heterogeneity across studies under the homozygote and recessive models . Conversely, significant heterogeneity across studies was reported under the heterozygote , dominant , and allelic  models. The sensitivity analysis found that the study by Janik-Papis et al. [p = 0.02; I2 = 67%), dominant , and allelic  models. Particularly, a significant association with AMD was confirmed under an allelic model .For the rs833061 polymorphism, the s et al.  affectedp < 0.001; I2 = 78%) and allelic  models. When the study by Janik-Papis et al. [p = 0.03; I2 = 64%) and allelic  models. However, no significant association was confirmed between the rs833061 polymorphism and wet AMD under both genetic models.In the stratified analysis of wet AMD, significant heterogeneity across studies was reported under the dominant . When the study by Janik-Papis et al. [I2 = 24% (p = 0.27) but no significant association was confirmed between the rs833061 polymorphism and AMD, in Caucasians, under the allelic model . Considering both populations, no significant heterogeneity across studies was reported in any of the other genetic models .In the subgroup of Caucasians, the s et al.  study wap = 0.13). Conversely, significant heterogeneity across studies was reported under the homozygote , dominant , recessive , and allelic  models. The sensitivity analysis found that the study by Almeida et al. [p = 0.10; I2 = 57%) and recessive  models. With regard to the dominant and allelic models, the sensitivity analysis found that the study by Churchill et al. [p = 0.47; I2 = 0%) and allelic  models. Particularly, a significant association with AMD was confirmed under the allelic model .For the rs1413711 polymorphism, no significant heterogeneity across studies was reported under the heterozygote model , dominant , and allelic  models. When the study by Churchill et al. [p = 0.08; I2 = 59%), dominant , and allelic  models. However, no significant association was reported between the rs833061 polymorphism and wet AMD under these genetic models.In the stratified analysis of wet AMD, significant heterogeneity across studies was reported under the homozygote .In the subgroup analysis by ethnicity, no significant heterogeneity across studies was reported in any of the genetic models . Conversely, significant heterogeneity across studies was reported under the homozygote model . The sensitivity analysis found that the study by Tian et al. [p = 0.31; I2 = 16%) and a significant association with AMD was showed under the homozygote model .For the rs3025039 polymorphism, the n et al.  affectedp = 0.004; I2 = 78%), heterozygote , recessive , dominant , and allelic  models. Particularly, when the study by Tian et al. [p = 0.17; I2 = 44%) and a significant association was reported between the rs833061 polymorphism and exudative form of AMD .In the stratified analysis of wet AMD, significant heterogeneity across studies was reported under the homozygote .In the subgroup analysis by ethnicity, no significant heterogeneity across studies was reported in any of the genetic models . Conversely, significant heterogeneity across studies was reported under the recessive model . The sensitivity analysis found that the study by Lin et al. [p = 0.50; I2 = 0%) but no significant association with AMD was showed under the recessive model.For the rs2010963 polymorphism, the p values > 0.1).In the subgroup analyses by subtypes of AMD and ethnicity, no significant heterogeneity across studies was reported in any of the genetic models . VEGF polymorphisms that can contribute to the risk of AMD.VEGF is a naturally occurring growth factor selective for endothelial cells that regulates angiogenesis and vascular permeability and plays a leading role in the retinal tissue of AMD, particularly wet AMD. In developed countries, this exudative type of AMD, characterized by the formation of subretinal choroidal neovascularization (CNV), is the major cause of severe vision loss and blindness in the elderly , 43. The VEGF polymorphisms in the AMD development, remain ambiguous, probably due to differences in patient selection, sample size, and genetic and environmental factors. These conflicting findings were summarized by three previous meta-analyses [ VEGF polymorphisms are associated with increased or decreased risk of AMD, particularly wet AMD [Results by several published studies, evaluating the role ofanalyses –23. The analyses . Furtheranalyses . Finally VEGF polymorphisms on risk of AMD.Our study, critically reviewing twelve studies, reports results of a more comprehensive meta-analysis and provides a stratified analysis for the exudative form of AMD. The results, by different genetic models, suggested the overall effects of VEGF −460 T/C polymorphism (rs833061), located in the promoter region, is associated with a decrease in the promoter activity of the gene [The T allele of thethe gene . The prethe gene , did notthe gene  was omitResults by the stratified analysis for wet AMD reported that C allele carriers had an increased risk of the exudative form of AMD under the homozygote, dominant, and allelic models. Though, these findings, under the dominant and the allelic models, were not confirmed by the sensitivity analysis.For the +674C/T polymorphism (rs1413711), located in intron 1 of the VEGF gene, the meta-analysis did not show an association with the overall AMD risk in any of the genetic models. The sensitivity analysis found that the study by Janik-Papis et al.  was a soAlthough the meta-analysis did not show an association between the +405G/C polymorphism (rs2010963) and the risk of AMD, the study by Tian et al.  was idenWith regard to the +936C/T(rs3025039) polymorphism, the present study indicated that it is not a risk factor for both overall AMD and wet AMD, in any of the genetic models.The present study has some limitations. First, the number of studies included in the meta-analysis is modest and some relevant articles were excluded due to insufficient data. Accordingly, the polymorphisms rs833061 and rs3025039 were analysed in six studies, the polymorphism rs1413711 was analyzed in four studies, and the polymorphism rs2010963 was analyzed in three studies. Second, the heterogeneity across studies, which existed for some polymorphisms, must be considered and, although the random effects model and the sensitivity analysis were appropriately performed, the pooled estimates should be interpreted with caution. Finally, since AMD is a complex disorder with sociodemographic, environmental, and genetic risk factors [ VEGF polymorphisms in the risk of AMD. Significant associations with AMD risk were showed for rs833061, rs1413711, and rs3025039 VEGF polymorphisms but not for rs2010963. However, given the above-mentioned limitations, further studies are needed to better clarify the effect of genetic susceptibility in the development of AMD.The present systematic review and the updated meta-analysis constitute a useful guide for readers to study AMD and add new evidence to the growing literature on the role of four common


Proper nutrition is important for overall health, and it reduces healthcare costs associated with malnutrition. Many studies have investigated vitamin D deficiency and its role in gestational diabetes and controversial data have reported. A comprehensive consideration of articles in this field provides the possibility of a general study of this relationship. This meta-analysis is an evaluation of the relationship between vitamin D deficiency and gestational diabetes.Different databases  were searched for studies and eligible English articles published before February 2017 that have reported the risk of gestational diabetes in relation to vitamin D deficiency. This relationship was measured using odds ratios (ORs) with a confidence interval (CI) of 95%. The influence of each study was measured through sensitivity analysis. Funnel plots, Egger regression tests, and the Begg–Mazumdar correlation test were used to determine bias or publication bias. STATA (version 11.2) was used for all analyses.p < 0.001). The serum level of 25(OH)D is less meaningful in people with gestational diabetes than in those who have normal glucose tolerance. Subgroup analysis showed that the results concerning this association may vary with study design but do not change with country of origin.Twenty-six studies were selected as eligible for this research and included in the final analysis. In general, vitamin D deficiency among mothers may be related to an increased risk of gestational diabetes (OR = 1.18; 95% CI, 1.01–1.35; Some evidence has shown that vitamin D deficiency may increase the risk of gestational diabetes. Gestational diabetes mellitus (GDM), a frequent disorder in pregnancy, is defined as any degree of glucose intolerance that first develops or is first recognized during pregnancy , 2. The It has been hypothesized that vitamin D deficiency is a risk factor associated with GDM. The need for vitamin D is higher during specific stages of life, including during the rapid growth of the fetus in the embryonic stage, infancy, the early stages of childhood, puberty, and pregnancy . In addiRecent evidence suggests that vitamin D receptors are expressed in many different cells, including those that influence the regulation of glucose metabolism such as muscle and pancreatic beta cells . VitaminThe association of vitamin D with the risk of gestational diabetes has long been of interest to researchers, and many studies in a broad variety of populations have been conducted on this topic, yielding different outcomes. Some studies have showed a meaningful relationship between vitamin D status and risk of gestational diabetes whereas others have reported controversial data. With this large body of literature, an overall estimation of its association is important, as is indicated by the breadth and quality of these studies. Recently, several meta-analyses have explored the existing data on the association of serum vitamin D levels with the risk of GDM and have reported that such association indeed exists –21. DespThis meta-analysis was done on the basis of the PRISAMA statement about meta-analyses (check list S1) . Two resFor further review, abstracts and titles were screened by two reviewers. Since every screened study was included in this meta-analysis, the researchers attempted to evaluate the relationship between vitamin D and the risk of GDM; determine whether it is accomplished among pregnant women without illness or as a chronic complication; use blood samples for lab tests; compare women with gestational diabetes and women with a normal glucose tolerance (NGT); report an estimation of effect  to compare adequate and inadequate vitamin D values. Some studies were out of the realm of the current study, which focused on pregnant women with chronic illness, and thus were excluded. Studies on non-human creatures , those published in languages other than English, those that were meta-analyses or systematic considerations, and those that presented insufficient data or were duplicate publications were also excluded.Data and information on the basis of a standard protocol were excluded after the determination was made for eligible articles. The data collected included the name of first writer, publication year, sample size, age group, present situation, study design, evaluation of the vitamin D status, and the effect estimation with a CI of 95%. In cases which needed more information, the articles’ writers were contacted for supplementary data or further elucidation. Information from articles by two specialists who worked independently was considered, and their findings and results were compared later. Disagreements about study eligibility were resolved by group discussion. The information and data were entered in a standard data extraction form and finally into Microsoft Excel.Q and I2 Cochran statistics. Wherever the results of studies were heterogeneous, a random effects model was used in the meta-analysis. Subgroup analyses were carried out in search of possible heterogeneous causes. Integrated estimations and the related CI of 95%were evaluated using forest plots as visuals. Bias or publication bias was evaluated as quality using funnel plots, Egger regressions, and the Begg–Mazumdar correlation test.Reported effect estimation (OR and RR) with a CI of 95% was used as the measuring standard of the relation between vitamin D and the risk of gestational diabetes. The possible estimation of risk was maximized to evaluate the variance between some methods and levels which control studies of potential confounding factors. Statistical heterogeneity was evaluated in studies using p < 0/05 were considered as valid for heterogeneity tests. For statistical analyses, software R (version 3.2.1) and STATA (version 11.1) software were used. All statistical tests were two sided.Values of In the initial electronic search, 268 potential articles were identified; a manual search of the bibliographies and reference lists of these articles identified 12 additional articles. Altogether, 280 articles were identified through the literature search. After the initial screening of abstracts and titles, 217 articles were excluded based on the inclusion criteria (49 articles based on unrelated titles and 168 based on unrelated abstracts) and 63 articles remained for full text review. In a secondary screening and after a full-text review, another 37 articles were excluded; 4 studies were not in a population of pregnant women, 8 studies were in pregnant women with chronic diseases or conditions, 6 studies did not obtain blood samples for testing, 3 studies reported on biological mechanisms of vitamin D metabolites, 7 studies had no outcome data, and 6 studies collected insufficient data. Two studies were non-human  studies, and one was not published in English. Twenty-six studies were selected for the final analysis after these exclusions  .Studies reporting OR were collected to determine the association between vitamin D deficiency and gestational diabetes. Seven studies showed a meaningful relationship between vitamin D status and risk of gestational diabetes, and 19 other cases showed no significant relationship. Thus, the present meta-analysis used a random effects model to show that vitamin D deficiency is related to an increased risk of GDM  Figure . The revI2 = 68.8%, p < 0.001)  (CI at 95% = −0.39 to −0.14), which indicates that heterogeneity was significant D was significantly lower in women with gestational diabetes than in women with NGT. Most of the reviewed studies reported that pregnant women with GDM have lower vitamin D values than the compared group , 43, 44.It is important to note that the relationship between the mother’s vitamin D status and the risk of gestational diabetes can be affected by various factors. Some possible contributing factors are mother’s age and body mass index (BMI). Different studies have shown a correlation between gestational diabetes and a mother’s high age, high BMI, and reduced physical activity , 48. ObeThe consumption of pre-natal multivitamins or nutritional supplements is one of the factors that can affect the relationship between maternal vitamin D status and the risk of developing gestational diabetes. A systematic study and meta-analysis showed that daily vitamin D supplementation  has protective effects on maternal health outcomes before and after childbirth and/or on the newborn . The resOther contributing factors or possibly risky factors may be exposure to sunlight, which has an intensive relationship with geographical situation, season, and weather; skin color and the use of sunscreen; and even air pollution . Recent Other factors can also be related to the risk of gestational diabetes, including smoking, alcohol consumption, weight gain associated with pregnancy, and social and economic status. The contribution of these confounding factors to the risk of GDM can go some distance in explaining the conflicting results among different studies.In the current study, subgroup analysis based on the research project of various studies in the meta-analysis showed that inadequate vitamin D is correlated with an increased risk of gestational diabetes if the studies did not use a cross sectional and cohort design. Since many of the studies reviewed in the present work had a case–control design, this may have led to overestimation of the effect size of the association. Similarly, previous meta-analyses confirmed the current results in this regard, showing that study design can be one explanation for heterogeneity among studies included in a meta-analysis –21.Another significant point is that entrance studies had used different methods and standards to recognize gestational diabetes. On the other hand, serum level 25(OH)D was measured in different quarterly courses, the definitions of deficiency or inefficiency of 25(OH)D were different, and different techniques were used. Some studies suggest that the method used to quantify 25(OH)D levels and the diagnosis of gestational diabetes could be an important factor and influence the final results , 55.2D3 to vitamin D receiver of β cells of the pancreas. Glucose stimulates the expression of insulin receptors and the 25-hydroxyvitamin D-1-α-hydroxylase enzyme and increases the insulin response for transfusion  The direct effect of vitamin D on the performance of β cells of the pancreas has a regulating effect on the rate of circulating glucose in the blood which is transmitted through the connection of 1,25(OH)nsfusion –58. (2) nsfusion . (3) Effnsfusion .This meta-analysis had several limitations. First, there were some differences among studies regarding the definition of the limits for vitamin D deficiency and the criteria for a diagnosis of GDM, which may be affected by accumulation. Second, the impact of some of the most important factors that may affect the relationship between vitamin D deficiency and gestational diabetes, including weight gain in pregnancy, skin color, and socioeconomic status, could not be evaluated due to inadequate data and information in some studies. Data on exposure to sunlight and vitamin D intake were not available. Third, the most adjusted OR was used in this meta-analysis. However, modulated models varied in different studies. Also, in some studies, the potential confounding factors could not be adjusted; so, the findings cannot be collected by adjusting the confounding factors. In addition, in meta-analysis sample size and SD are very important in combining the results of studies; this subject affected our study. Furthermore, the possible effects of publication bias inherent in any meta-analysis cannot be ruled out.The present meta-analysis shows that women with GDM had significantly less serum vitamin D than women who have NGT. The current finding shows that low levels of 25(OH)D may be considered a risk factor during gestation. Also, the results showed that there is a mild positive association between maternal vitamin D deficiency and GDM and suggested softening the reported conclusion in previous meta-analysis. In conclusion, our results show that vitamin D deficiency may increase the risk of gestational diabetes. Based on different research projects and the heterogeneity among various studies, it is not logical to make a definitive conclusion. More clinical trials and studies are needed to determine the effects of vitamin D supplements on the prevention of gestational diabetes. Based on these findings, healthcare providers should encourage pregnant women to follow the guidelines for the daily intake of vitamin D.This study has a code of ethics by the Ethics Committee of Ilam University of Medical Sciences (ir.medilam.rec.1395.102).MA and SM designed the study and participated in data collection; KS, AM, and ES performed the meta-analysis; AM, SFM, ES, and SM participated in drafting the paper; KS and ES critically revised the paper; all authors provided data analysis, read, and approved the manuscript.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


P=6.16E-21, odds ratio (OR) =0.92, 95% confidence interval (CI) 0.91-0.94). In order to validate the effect of rs9929218 variant on CDH1 expression, we further performed a functional analysis using two large-scale expression datasets. We identified significant regulation relation between rs9929218 variant and the expression of CDH1, ZFP90, RP11-354M1.2 and MCOLN2 by both cis-effect and trans-effect. In summary, our analysis highlights significant association between rs9929218 polymorphism and CRC susceptibility.Colorectal cancer (CRC) is the third most common cancer. Large-scale genome-wide association studies (GWAS) have been performed and reported some novel CRC susceptibility variants in European ancestry including the CDH1 rs9929218. Following GWAS and candidate studies evaluated the association between the CDH1 rs9929218 polymorphism and CRC in European, Asian and American populations. However, these studies reported inconsistent associations. Evidence shows that rs9929218 may regulate different gene expressions in different human tissues. Here, we reevaluated this association using large-scale samples from 16 studies (n=131768) using a meta-analysis method. In heterogeneity test, we did not identify significant heterogeneity among these studies. Meta-analysis using fixed effect model showed significant association between rs9929218 and CRC ( Colorectal cancer (CRC) is the second leading cause of cancer death in developed countries . It is aP = 8.10E-10), 16q22.1 ; P = 1.20E-8), 19q13.1 ; P = 4.60E-09) and 20p12.3  [In 2008, Houlston et al. conducted a meta-analysis of two CRC GWAS including 13,315 individuals in discovery dataset and 27,418 subjects from eight independent case-control series in replication dataset . They idFollowed the original CRC GWAS, several GWAS and candidate variant studies also investigate the association between rs9929218 and CRC in European, Asian and American populations –25. HoweWe selected 57 and 140 articles from PubMed and Google Scholar databases. Using the inclusion criteria and exclusion criteria, 11 articles including 16 independent studies were finally selected for our following analysis. More detailed information about the inclusion or exclusion of selected studies was described in Figure I2 statistic, respectively. Using Cochran's Q statistic, we identified no significant heterogeneity among these 16 studies with Cochran's Q statistic = 17.62, and degrees of freedom = 15, and P=0.2829. We identified I2 = 14.9%, which indicates no statistically significant heterogeneity. We further performed a subgroup heterogeneity test in Asian, American and European populations, respectively. In the end, we found no significant genetic heterogeneity in Asian population with I2 = 0 and P=0.70, American population with I2 = 29% and P=0.19, and European population I2 = 11.2 and P=0.32.We evaluated the genetic heterogeneity using Cochran's Q statistic and P=6.16E-21, OR=0.92, 95% CI 0.91-0.94). Detailed results are described in Figure Considering no significant heterogeneity in all the selected studies, we calculated the overall OR by the fixed effect model. The meta-analysis results showed significant association between rs9929218 and CRC , American population , and European population .We further performed a subgroup meta-analysis in Asian, American and European populations, respectively. The results further supported the significant association between rs9929218 and CRC in Asian population  can significantly regulate ZFP90 expression in the whole blood, Testis and Muscle-Skeletal tissues with P=6.60E-14, P=5.90E-06, and P=1.60E-05 by a cis-regulation. Meanwhile, rs9929218 can also regulate CDH1 expression in Nerve-Tibial and Spleen tissues with P=2.80E-11 and P=1.20E-06, respectively  can significantly regulate ZFP90 expression with I2statistic, we did not identify significant heterogeneity among these 16 studies. Using a fixed effect model for meta-analysis to calculate the overall OR, we identified significant association between rs9929218 and CRC . The sensitivity analysis and publication bias further supported our results to be robust. After excluding the two large-scale studies from Houlston [GWAS reported rs9929218 to be significantly associated with CRC in European ancestry . FollowiHoulston , we stilcis-effect and trans-effect. Recently, Closa et al. analyzed the association between genotypes for 26 GWAS SNPs and the expression of genes within a 2 Mb region (cis-eQTLs) using Affymetrix Human Genome U219 expression arrays with 47 samples from healthy colonic mucosa and 97 samples from normal mucosa adjacent to colon cancer [P=3.90E-03 [P=5.50E-03, P=6.70E-03 and P=1.70E-03, respectively, but not in normal tissue eQTL analysis with P=0.59, P=0.54 and P=0.27, respectively [Evidence shows that eQTL may modify gene expression and influence risk for human diseases –29. In oIt is reported that the rs9929218 variant and rs1862748 are in linkage disequilibrium (LD) with rs28626308, which results in a non-synonymous change in RHPN2 protein . Their fHere, we reinvestigated the association between rs9929218 and CRC using an additive model as descried in Figure We searched the PubMed database to select all possible studies with the combined key words including ‘CDH1′, ‘rs9929218′, ‘polymorphism’, or ‘variant’ and ‘colorectal cancer’, ‘colon cancer’, ‘rectal cancer’, or ‘bowel cancer’. We also used the Google Scholar to query the articles citing CRC GWAS and select all associated publications. There was no language and sample limitation, and the literature search was updated on October 28, 2015. This search was performed by two independent individuals to identify potential publications meeting our inclusion criteria (see below), with differences resolved by discussion with a third reviewer. More detailed information about the search strategy is described in previous studies –35.The following inclusion criteria were used to select all potential publications for the meta-analysis: (1) use a case-control design; (2) evaluate the association between rs9929218 and CRC; (3) provide odds ratio (OR) with 95% confidence interval (CI) for allele model (see below); or (4) provide sufficient data to calculate the OR and 95% CI for allele model; For any publication, it is excluded from the meta-analysis, if (1) not for cancer research; (2) not use a case-control design including case population; case reports, conference abstract, reviews; (3) not provide OR with 95% CI for allele model (see below); or (4) not provide sufficient data to calculate the OR and 95% CI for allele model; More detailed information is described in previous studies –35.Two authors reviewed and extracted information from all eligible publications independently, according to the inclusion and exclusion criteria listed above. The following items were extracted including (1) the name of the first author; (2) the year of publication; (3) the population and ethnicity; (4) the numbers of CRC cases and controls; (5) the OR with 95% CI or (6) to calculate the OR and 95% CI; If there is a conflict, both reviewers discuss with third reviewer and reach an agreement. For rs9929218 polymorphism A/G alleles, we selected the additive genetic model: A allele versus G allele for further meta-analysis –40.χ2 distribution with k-1 degrees of freedom  [P < 0.01) indicates heterogeneity among selected studies. I2 is a measure of heterogeneity and a statistic that indicates the percentage of variance in a meta-analysis that is attributable to study heterogeneity [I2 > 50% indicates statistically significant heterogeneity.Genetic heterogeneity is evaluated using two methods including Cochran's Q-statistic and tatistic –46. Cochnalysis) –46. A siogeneity –46. The ogeneity –46. The P < 0.01) indicates significant association between rs9929218 and CRC. Using sensitivity analysis, each study is omitted at a time to assess the influence of single study on the association between rrs9929218 and CRC. A funnel plot proposed by Egger et al. is used to investigate potential publication bias [P < 0.01 indicating that there was a significant publication bias [Here, we used two kinds of meta-analysis models including the fixed effect model (Mantel-Haenszel) and random-effect model (DerSimonian-Laird) to calculate the pooled OR. If there is no significant heterogeneity among the included studies, the pooled OR is calculated by the fixed effect model; otherwise the OR is calculated by random-effect model –46. Z teion bias –46. A liion bias . All staion bias –46.To validate the effect of rs9929218 variant on CDH1 mRNA expression level, we performed an expression quantitative trait loci (eQTL) analysis using a large-scale eQTL dataset, which is from a meta-analysis in non-transformed peripheral blood samples from 5,311 individuals with replication in 2,775 individuals .http://www.gtexportal.org/home/), which provides a scientific resource to study human gene expression and regulation and its relationship to genetic variation [We also investigate the effect of rs9929218 variant on CDH1 mRNA expression using the Genotype-Tissue Expression (GTEx) project (ariation . Here, wariation .