A proportional hazards measure is suggested in the context of analyzing SROC curves that arise in the meta–analysis of diagnostic studies. The measure can be motivated as a special model: the Lehmann model for ROC curves. The Lehmann model involves study–specific sensitivities and specificities and a diagnostic accuracy parameter which connects the two.A study–specific model is estimated for each study, and the resulting study-specific estimate of diagnostic accuracy is taken as an outcome measure for a mixed model with a random study effect and other study-level covariates as fixed effects. The variance component model becomes estimable by deriving within-study variances, depending on the outcome measure of choice. In contrast to existing approaches – usually of bivariate nature for the outcome measures – the suggested approach is univariate and, hence, allows easily the application of conventional mixed modelling.proc mixed allow the fitting of mixed models for meta-analytic data from diagnostic studies. The methodology is illustrated with several meta–analytic diagnostic data sets, including a meta–analysis of the Mini–Mental State Examination as a diagnostic device for dementia and mild cognitive impairment.Some simple modifications in the SAS procedure The proposed methodology allows us to embed the meta-analysis of diagnostic studies into the well–developed area of mixed modelling. Different outcome measures, specifically from the perspective of whether a local or a global measure of diagnostic accuracy should be applied, are discussed as well. In particular, variation in cut-off value is discussed together with recommendations on choosing the best cut-off value. We also show how this problem can be addressed with the proposed methodology. The parameter β allows for an association between test accuracy and test threshold. When β=0, αi is estimated by Di and θi is estimated by Si/2, where Di and Si are as for the Littenberg–Moses model. Furthermore, to account for between-study variation, a random effect is assumed for θi and αi being independent. As an alternative, a bivariate normal random-effects meta–analysis has been suggested by van Houwelingen et al. Using proportional hazards (PH) measure. In Figure We suggest using the measure One of the features of the SROC approach is that it incorporates the cut-off value variation in a natural way; hence a measure modelling an ROC curve is favorable. We suggest the PH measure based upon the Lehman family in the following way: q, (1−q)θ is also feasible as long as θ is positive. Note that (1) is defined for all values of p∈ and q∈  whereas p∈ and q∈. Population values of sensitivity and specificity of 1 are rarely realistic, although observed values of 1 for sensitivity and specificity do occur in samples. This can be coped with by using an appropriate smoothing constant such as estimating specificity as (ni−1)/ni when xi=ni and sensitivity as (mi−1)/mi if yi=mi.This model was suggested by Le  for the θ is called the proportional hazards measure. By taking logarithms on both sides of (1) we achieve In Figure t. Hence the name proportional hazards model, which was suggested in a paper by Le 2Var(X) assuming that the variance Var(X) of X is known. Applying this δ−method twice gives and apply the and i-th study is provided as so that the within study variance for the We acknowledge that the above are estimates of the variances of the diagnostic accuracy estimates, but are used as if they were the true variances.Some important cases. If there are no further covariates, two important models are easily identified as special cases of (7). One is the fixed effects model random effects model and the other is the which have gained some popularity in the meta-analytic literature.We illustrate the approach with an example and revisit a meta–analysis by Mitchell  on the dproc mixed from the SAS software, version 9.2 for Windows [condition which takes the value 1 if the study concerns MCI as condition and 0 if the study is on dementia. Since we have fixed within-study variances, we need to tell proc mixed to incorporate this appropriately; this can be accomplished by using a weight, random option induces a random effect (here study) with associated variance component τ2, which is estimated. However, SAS proc mixed will automatically fit a within-study variance component (on top of the provided variances). To circumvent this mechanism, the option parms (1) (1) /hold=2 is used where the term hold=2 fixes the second variance component, corresponding to the within-study variance multiplier, to one. Note that the random effect modelling between-study variation is described by a free variance parameter, τ2. For this a starting value needs to be given: we have τ2=1, although other choices are possible, e.g. τ2=0, corresponding to the case of no heterogeneity between studies.We use  Windows , for theThe results of the analysis are provided in Table The inference is based here on a procedure called the Wald test. The estimated parameter value is divided by its estimated standard error, and the result is given in column four in Table It is trivial to construct the associated SROC curves from Table 0:τ2=0. In this case, the asymptotic null distribution of the likelihood ratio test statistic is no longer χ2 with 1 df but rather a mixture of a two-mass distribution giving equal weights 0.5 to the one-point mass distribution at 0 and a χ2 with 1 df [Note that the likelihood ratio test as well as the Wald test need modification in situations where the null hypothesis is part of the boundary of the alternative such as when testing H ith 1 df . Practicet al.[The MOOD module of the Patient Health Questionnaire (PHQ-9) has been developed to screen and to diagnose patients in primary care with depressive disorders. The instrument consists of 9 questions, each scored from 0 to 3 points with a total score ranging from 0 to 27. In a meta–analysis of the diagnostic accuracy of MOOD, Wittkampf et al. includedcut-off, which is zero when the summary score with a cut-off value of 10 is used and one otherwise. The results are presented in Table cut-off level “summary score” is associated with a higher diagnostic accuracy, although, as seen from the Wald statistics provided in column four of Table cut-off is not significant, since the concept of the SROC is designed to accommodate the cut-off value variation. We will take up this point in the next section.The presence or absence of a cut-off value effect is now more formally investigated using a covariate globally, in the sense that cut-off value variation will not necessarily effect the estimate of the SROC curve. The situation is illustrated in Figure We have focussed on the PH measure so far, as it provides an appropriate measure for comparing SROC curves θ, hence, the PH measure logθ is not the best measure to discriminate different cut-off values. This is not surprising, since the SROC curve is a concept designed for assessing the diagnostic accuracy of a diagnostic test globally, in the sense that it adjusts for different cut-off values. Hence, a measure that assesses local performance of the diagnostic is needed. Assuming that every cut-off value used in the meta–analysis is clinically meaningful, we suggest use of the (squared) Euclidean distance to the upper left corner  of the ROC diagram as a more meaningful measure to compare cut-off values: Evidently, different cut-off values are associated with the same value of logwhere δ-method once more, to obtain Since we have changed the criterion, we need to determine the associated within-study variances. This can be accomplished easily, using the where we have ignored study indexes for the the sake of simplicity. Using this criterion, we see in Figure Evidently, both criteria lead to the same conclusion, namely that using the summary score with a cut-off value of 10 leads to the higher diagnostic accuracy . It might also be worthwhile looking at the results of the likelihood ratio test: for the PH-measure as the outcome variable, the likelihood ratio test provides a value of 1.5; for the Euclidean distance, the value of the likelihood ratio test is 1.7, confirming the non-significance of the effect. Nevertheless, the analysis shows that the cut-off value of 10 provides the higher diagnsotic accuracy.et al.[This case study provides an example where the use of a global or local criterion leads to a different conclusion. Magnetic resonance spectroscopy has the ability to discriminate between prostate cancer and benign prostatic hyperplasia, based on reduced citrate and elevated choline in the cancerous region. The diagnostic test works on a voxel of signal intensity ratios of (choline+creatine)/citrate. Two cut-off points are in use: <0.75 and <0.86. The results collected in a meta–analysis by Wang et al. include The fixed effects parts of the mixed model analysis, using the global PH measure and the local Euclidean measure as criteria, are presented in Table P(T=1|D=1)/P(T=1|D=0) or p/(1−q). It is different to the PH measure in that the ratio is taken on the log-scale: θ= logp/ log(1−q). Furthermore, if re-expressed as models, the positive-likelihood ratio corresponds to p=θ′(1−q), a straight line with no intercept, whereas the the PH measure corresponds to p=(1−q)θ, a straight line on the log-scale with no intercept. The positive likelihood ratio is a natural measure since it transfers the concept of relative risk (risk of a positive test in the diseased group to the risk of a positive test in the non-diseased group) to the diagnostic setting. However, it is less suitable as an (S)ROC model since it does not provide a function which connects the lower left vertex with the upper right vertex in the ROC diagram .Another frequently used diagnostic measure is the positive likelihood ratio, defined as the ratio of sensitivity to false positive rate The approach presented here is attractive since it is based on a simple measure of diagnostic accuracy per study, namely the ratio of log-sensitivity to log-false-positive rate. It also embeds the diagnostic meta-analysis problem into the well-known and much used mixed model setting. In particular, the analysis of effects of observed covariates on the diagnostic accuracy can easily be incorporated.Controversies in the meta–analysis of diagnostic studies usually focus on comparability of studies. Study types might be case–control, cohort, cross–sectional or other. Studies might differ in the gold standard, severity of disease, or in the application of the diagnostic test. Patient populations might differ across studies, as might the cut-off value (defining positivity of the diagnostic test). All these different aspects, if observed, can be easily incorporated and analyzed as fixed effects in the special mixed model suggested here.τ2). This might allow us to construct a measure of relative residual heterogeneity, which might help to assess how trustworthy the results of a given meta-analysis may be. This will be investigated in future research.The occurrence of heterogeneity in the meta-analysis of diagnostic studies is more the rule than the exception; it is thus important to quantify the heterogeneity across studies due to the different sources. The approach provided here offers a more detailed investigation of heterogeneity according to the various observed sources and a residual heterogeneity .Twenty-six studies were included in their meta-analysis with a wide range of non-evaluable results from 0 to 43. Using the approach suggested here, it would be very easy to analyze the effect of non-evaluable results on the diagnostic accuracy by including the amount of non-evaluable results per study as a fixed effect in the proposed mixed model.The authors declare that they have no competing interests.SC carried out all statistical and computing analysis. WB collected and prepared all meta-analytic data sets. HH conceived the theoretical modelling work and DB drafted the idea of the approach and finalized the manuscript. All authors read and approved the final manuscript.The pre-publication history for this paper can be accessed here:http://www.biomedcentral.com/1471-2288/14/56/prepub


Begg = 0.296, PEgger = 0.161) was found. Overall, there was statistically significant association for recessive genetic model . Moreover, the positive result was confirmed using trial sequential analysis even only three original studies. For allele model, there was also statistically significant association . Meanwhile, A allele as a risk factor turned out to be true positive by trial sequential analysis. In a word, this meta-analysis suggested that the single nucleotide polymorphism 12 definitely increase the risk of hypospadias.To investigate the association between single nucleotide polymorphism 12 in estrogen receptor α gene and hypospadias, four databases  were electronically searched by 2nd November 2015. Finally, four studies were included for our meta-analysis, involving 1379 cases and 1648 controls. A quality assessment was performed using the Newcastle–Ottawa Scale of case–control study. Meta-analysis and publication bias measuring were all done by Stata 12.0. No significant publication bias (P Hypospadias is the second most common congenital malformations of the male external genitalia among male newborns, in which the urinary opening is located along the ventral side of the penis. The prevalence of hypospadias varies widely between countries and populations, ranging from 3.9 to 43.2 cases per 10,000 births . ESR1 is on chromosome 6, which expresses in most cells of the male urethra  related gene with hypospadias is a new research hotspot, especially estrogen receptor α gene (ESR1 and hypospadias by 2nd November 2015. Searching terms were: (‘estrogen receptor’ OR ‘ESR1’ OR ‘ESR1 single-nucleotide polymorphism12’ OR ‘rs6932902’) AND . In additional, we checked the reference lists of retrieved reviews to identify more pertinent studies.Four databases  were electronically searched to retrieve studies on the associations between SNP12 in ESR1 polymorphism and hypospadias; (2) the ESR1 polymorphism was tested in the study; (3) the papers identified the odds ratio (OR) and the 95 % confidence interval (CI) or other information that can help to infer the target data; (4) the study design was a case–control study; (5) when multiple publications reported on the same or overlapping data, the most recent article or the article based on the largest study population was selected; (6) the publication language was English.Titles and abstracts of all relevant papers were reviewed firstly. Then, full-texts were reviewed as a second screening. The studies were considered eligible if they met all of the following criteria: (1) the study explored the relationship between Studies met any of the following exclusion criteria were excluded: (1) researches based on animals or cells rather than general population; (2) reviews, editorials, meeting abstracts, and commentaries; (3) articles with no target data or no relevant outcomes.Two reviewers (Changkai Deng and Rong Dai) independently searched the literatures and selected the papers; then, they extracted the relevant data in accordance with the pre-formed data extraction form. Disagreements were solved by discussion and a third party (Xuliang Li and Feng Liu) was involved when necessary. The basic information was extracted from each article: first author, year of publication, country where study was conducted, ethnicity of subjects, sample size of case and control groups, demographic characteristics of case and control groups (age), study design, method of genotyping assay, deviation from Hardy–Weinberg Equilibrium (HWE).Quality assessment was conducted for each article by using the Newcastle–Ottawa Scale of case–control study. The quality of the studies was evaluated by examining three items: selection of case and controls, comparability of groups and ascertainment of exposure. Studies were graded on an ordinal star scoring scale with higher scores representing studies of higher quality. A study can be awarded a maximum of one star for each numbered item within the selection and exposure categories and a maximum of two stars can be given for the comparability of the two groups. The quality of each study was graded as high quality (7–9) and low quality (1–4). For the present studies, Changkai Deng and Rong Dai scored each included study independently.ESR1 and hypospadias. Heterogeneity among included studies was checked by Chi square-based Q test and I2 test. If the data showed low heterogeneity , Mantel–Haenszel fix effect model was used, otherwise DerSimonian–Laird random effect model was used. The pooled ORs were performed for the allele contrast (A vs. G) and recessive genetic model (AA vs. GA + GG). HWE was tested for included studies using an online HWE calculation tool and publication bias was assessed by Egger’s linear regression test. Data were analyzed using Stata 12.0. A P value of 0.05 for any test or model was considered to be statistically significant unless otherwise specified.ORs with their 95 % CIs for alleles and genotypes were used to assess the strength of association between the SNP12 in Potential publication bias was assessed by the Begg rank correlation test and Egger linear regression test using the software of Stata 12.0.TSA can adjust the threshold for statistical significance according to the quantified strength of evidence and the impact of multiplicity. A meta-analysis may result in type I errors and type II errors if data are sparse or if there is repeated testing for significance when new trials are added .The results of the Begg’s rank correlation test and Egger’s linear regression test supported the conclusion of no significant publication bias  during fetal development may result in formation of abnormal male urethra  is 3.45  and allele contrast (A vs. G) is 1.43 . The results indicate that there is significant association between SNP12 in In this meta-analysis, there are some limitations which may influence the findings. Firstly, we focused on individual SNP. Polymorphisms occur at low frequencies in many ethnic groups and are unlikely to explain the variation in response when considered alone. Secondly, our meta-analysis has not taken the possibility of SNP–SNP, gene–gene or gene–environment interactions into consideration.This meta-analysis suggested that the single nucleotide polymorphism 12 definitely increase the risk of hypospadias.


Previous studies have yielded controversial results related to the contribution of matrix metalloproteinase-2 (MMP-2) -1306 C/T and -735 C/T polymorphisms in the progression of coronary artery disease (CAD). This study aimed to provide strong evidence for the role of the 2 polymorphisms in genetic risk of CAD.The human case-control studies regarding the association of MMP-2 polymorphisms with CAD risk were systematically identified through online databases  and manual search. Inclusion criteria were defined for the eligible studies. The fixed-effects meta-analysis was performed to combine the values when homogeneity was indicated. Alternatively, the random-effects meta-analysis was utilized.A total of 2118 samples were analyzed in the meta-analysis of -1306 C/T. The odds ratio for the initially tested genetic model was 0.93 . The remaining comparisons similarly showed -1306 C/T genotypes were not significantly associated with the risk of CAD. We noted the same trend when data were retrained to myocardial infarction studies. Meta-analysis of -735 C/T suggested no clear association with the development of CAD.The results of the current work fail to support a significant involvement of MMP-2 -1306 C/T and -735 C/T polymorphisms in the risk of developing CAD. Activation of MMP is believed to accelerate plaque rupture and thrombus formation by impairing the fibrous cap and degrading extracellular matrix proteins.9 The polymorphic MMP-2 is located at chromosome 16q13 with 13 exons within its promoter region. MMP-2 single nucleotide polymorphisms (SNPs) enable alterations in transcriptional regulation and thereby affect the function of its enzyme.10 These functional SNPs also affect MMP-2 gene transcription in an allele-specific fashion and are thus considered potential candidate to evaluate the associations with atherogenesis, tumor initiation, invasion, and metastasis.11Matrix metalloproteinase-2 (MMP-2) is an important member of MMP family involved in several cellular and natural processes , and functions as a proteolytic enzyme catalyzing matrix proteins, basement membrane constituents in particular.12 Most strikingly, the -735 T allele even leads to elimination of the promoter activities.13 Due to the functional importance, several groups have associated the 2 polymorphisms with incident CAD, a heterogeneous and common vascular disorder that has caused a large number of deaths worldwide.6 The published data nevertheless fail to yield a consistent conclusion most likely because of the inadequate sample size.15 Herein, we aimed to determine whether or not the MMP-2 SNPs act as modifiable risk factors for CAD by use of a meta-analysis.There have been many polymorphisms widely studied in CAD community, including MMP-2 -1306 C/T and -735 C/T promoter SNPs. A notably decreased or increased enzyme activity has been linked to the -1306 T allele and the C allele, respectively.The study was approved by the ethics committee of general hospital of Beijing military command of PLA.Online databases  were used to seek the studies of the association between MMP-2 polymorphisms and CAD risk, without language restriction. The MeSH search terms and keywords included coronary artery disease, matrix metalloproteinase-2, polymorphism, polymorphisms, and their synonyms  and abbreviations. Additional data were obtained by scanning the reference lists of review articles and research articles that fulfilled all of the following inclusion criteria:First, the study must be published before the literature search was terminated (December 2013);Second, CAD or MI patients must be investigated and the control group comprised healthy individuals rather than non-CAD or non-MI patients to guarantee the accuracy of estimations;Third, at least 1 of the MMP-2 polymorphisms, either -1306 C/T or -735 C/T, was studied;Fourth, the original article must report ample genetic data by which the risk of CAD could be estimated;Fifth, there was no notable deviation from Hardy–Weinberg equilibrium (HWE) in control group.Any study violating the predesigned criteria was not considered in this meta-analysis.To minimize the possibility of biased results induced by erroneous data abstraction, 2 investigators independently extracted authors, country of origin, ethnicity, publication year, matching status, adjusted factors, type of disease (CAD or MI), and genotype frequency, using a standardized protocol. Discrepancies were resolved through discussion with the third investigator.16; conversely, the random-effects meta-analysis was employed.17 The degree of heterogeneity across studies was evaluated using the I2 statistic,18 with I2 >50% being considered large heterogeneity. In order to check if the combined estimates were stable, more specifically, whether the single studies influenced the results materially, we performed sensitivity analysis. HWE in controls was determined using the χ2 test. Both the funnel plots and Egger linear regression test were used to evaluate the publication bias in this meta-analysis.19Crude odds ratios and its 95% confidence intervals were pooled to evaluate the correlation between MMP-2 polymorphisms and CAD risk. We first used TT + CT versus CC genetic comparison, then TT versus CT + CC, TT versus CC, and CT versus CC. Pooled ORs were combined using the fixed-effects meta-analysis in the absence of substantial heterogeneityAll statistical data were analyzed via the Stata software . The significance threshold was set at P values below .05.23 . We therefore identified 6 eligible studies for this analysisAs displayed in Table Fixed-effects meta-analysis of 1059 CAD cases and 1059 healthy controls suggested that individuals carrying the TT + CT genotypes of MMP-2 -1306 C/T did not have higher or lower risk of CAD  .We used Begg funnel plot and Egger test to evaluate the publication bias. The studies of MMP-2 polymorphisms were symmetrically distributed within the funnel plots and the Egger test provided statistical evidence of little publication bias across the literature  (Figure 24 Although external factors such as changes in dietary patterns, an increased prevalence of smoking, and intense pressure are taken as contributory factors for the development of CAD, the research focus in recent years has shifted to the genetic contributions to this invasive heart disease. On the basis of 1059 patients with CAD and 1059 control subjects, we estimated the relationship of MMP-2 polymorphisms with CAD susceptibility by means of meta-analysis, a widely used quantitative approach that could provide strong evidence for SNP-malignancy associations. We found -1306 C/T genotypes were not significantly associated with CAD risk using the genetic models listed in Table CAD represents a major type of heart disease and has been a global health problem bedevilling an increasing number of people.26 A recent study has associated atherosclerotic plaques with enhanced MMP expression levels.27 Downregulation of MMP-2 expression has been reported in both acute and chronic phases of coronary disease in vitro.28 Available data also demonstrated elevated circulating levels of MMP-2 in patients with acute coronary syndrome compared to those with stable effort angina.30 In patients with acute MI, Hojo et al observed a gradually increased plasma MMP-2 levels and the increase reached to the maximum on day 21 after onset.31 According to the previous data, we can speculate that the polymorphisms, the commonest variations in human genomics, in the promoter region of predisposition genes, such as MMP-2, probably act as modifiable risk factors for the development and progression of CAD. This speculation fails to lend support to the findings in the current work, where we found no association of MMP-2 polymorphisms and CAD incidence. There are several possible reasons for the inconsistency. The MMP-2 promoter polymorphisms as low-penetrance sequence variations require a large-scale study to detect their slight or moderate effects on CAD risk. Further, the different methodologies applied in each of the published studies may decrease the precision of genotyping and hence leads to biased results in the single studies and this meta-analysis consequently.Earlier studies have established the contribution of MMP activity to angiopathy initiation by transferring vascular smooth muscle cells into the intimal space which provides microenvironment for cellular proliferation and plaque formation via the internal elastic lamina.22 Replication efforts for the association included Ghaderian et al, who found no statistically significant association between genetic polymorphisms of MMP-2 and risk of MI.15 No significant associations were later reported in a Turkish study.20 With respect to the 2 Chinese studies, one is for -1306 C/T and the other for -735 C/T, both of which failed to identify any significant relation with CAD.23 As most of the previously published studies have shown no involvement of the MMP-2 polymorphisms in modifying the risk of CAD or MI, we can infer that the genetic variants of interest are unlikely to play a major role in the disease concerned in this study.MMP-2 polymorphisms are of peculiar interest due to the functionally important role played in transcriptional activation activity of MMP-2 protein, including -1306 C/T and -735 C/T. A study of Mexican samples showed 2.05-fold increased risk of MI attributable to the -1306 CC genotype and this increase seemed more pronounced in obese or hypertensive subjects.14 It is interesting that a most recent study, also in a Mexican population, reported MMP-2 -1575G/A polymorphism, but not -1306 C/T and -735 C/T, was an independent risk factor involved in the risk of developing MI.In order to facilitate a better understanding of the effect estimations, several limitations of this analysis should be noted. First, our meta-analysis incorporated a limited number of studies, with each representing a small or moderate sample size, and this may result in false negatives as reported in the present work. Second, it is still unclear whether or not the MMP-2 polymorphisms modulate the risk of CAD or MI when data are restrained to some specific population, such as Mexican, Caucasian, and East Asian. Third, due to the considerable difference in adjustment confounders across the literature, we cannot evaluate the risk of CAD using an adjusted odds ratio.Nevertheless, we have directed much effort to literature search and combined all data published to date, suggesting the first meta-analysis of MMP-2 -1306 C/T and -735 C/T SNPs has provided the strongest evidence of the genetic associations than any previous study. Therefore, our results are relatively robust and may have some implications for future investigations.In conclusion, current evidence failed to support a significant association between -1306 C/T and -735 C/T polymorphisms at MMP-2 locus and the risk of CAD. Further research is clearly needed to determine the role of the 2 polymorphisms in the development of CAD and thus provides novel insights into the molecular mechanism that underlies genetic risk of the pervasive disease.


Some studies have investigated the effects of polymorphisms in the vascular endothelial growth factor (VEGF) gene on responsiveness to chemotherapy for colorectal cancer (CRC) and have shown inconclusive results.Eligible studies that assessed the associations between polymorphisms in the VEGF gene and response to chemotherapy in CRC were searched in the PubMed, Embase and Medline databases until November 2014. Odds ratios (OR) and 95% confidence intervals (CIs) were used to evaluate the associations, using Review Manager software, version 5.3. Stratified analysis was also conducted.In the overall analysis, a significant association with responsiveness to chemotherapy in CRC was identified in CC vs. CA of the VEGF -2578 C/A polymorphism  and in CC+CT vs. TT of the VEGF -460 C/T polymorphism . In subgroup analysis, a significant association was found in excluding anti-angiogenic agent subgroup in three comparison models of the VEGF -2578 C/A polymorphism and another three genetic models of the VEGF -460 C/T C/A polymorphism.CC vs. CA of the VEGF -2578 C/A polymorphism and CC+CT vs. TT of the VEGF -460 C/T polymorphism might be predictive factors of responsiveness to chemotherapy in CRC. However, single-nucleotide polymorphisms in the VEGF gene lacked sufficient predictive ability to determine whether patients with CRC should add anti-angiogenic agents to their chemotherapy regimens. Colorectal cancer (CRC) is one of the leading causes of death worldwide, and approximately 1 million people are diagnosed with CRC every year . It is aCurrently, chemotherapy is widely used in malignant tumors for significant improvements in overall survival (OS) and progression free survival (PFS) in patients –5. RegarThe VEGF gene is located on chromosome 6p21.3, and its coding region spans approximately 14 kilobases and consists of 8 exons –11. The CRC is a complicated disease affected by both genetic polymorphisms and environmental factors –15. VEGFSNPs in the VEGF gene, including -2578 C/A, -460C/T, +405G/C, and +936C/T, have been focused in the relationship of the gene with the response to chemotherapy in CRC –24. HoweIn our work, a meta-analysis of all published studies was performed to investigate whether VEGF polymorphisms were associated with responsiveness to chemotherapy in patients with CRC. Moreover, a subgroup analysis with regard to a combination of anti-angiogenic agents in chemotherapy regimens was also performed, to investigate whether SNPs in the VEGF gene could work as biomarkers to predict the outcomes of adding anti-angiogenic agents to chemotherapies for CRC. As far as we know, this was the first systemic review and meta-analysis that focused on the associations between VEGF gene polymorphisms and response to chemotherapy for CRC.All studies assessing the associations between polymorphisms in the VEGF gene and response to chemotherapy in CRC were retrieved via an exhaust search of databases, including PubMed, Embase and Medline. The bibliographic search was performed by two investigators using the following retrieval terms:  AND  OR "genetic polymorphism"[All Fields] OR "polymorphism"[All Fields])) AND ) AND  OR "colorectal neoplasms"[All Fields] OR  OR "colorectal cancer"[All Fields]. Other potentially eligible studies were found by manually searching relevant reviews and the included studies. All of the records were updated to November 2014. Only English-language articles were used for this meta-analysis.For inclusion in the meta-analysis, studies had to meet the following criteria: (1) studies assessing the associations between polymorphisms in the VEGF gene and response to chemotherapy in CRC; (2) independent prospective or retrospective association studies; and (3) studies providing detailed data to estimate odds ratios (ORs) and corresponding 95% confidence intervals (CIs). Studies were excluded if they were other types of original studies, such as reviews, meta-analyses and case reports. Moreover, studies were also not eligible for this meta-analysis if they lacked critical information.The data extraction was conducted independently by two investigators (L. Wang and S. Ji). Inter-researcher discrepancies were settled by discussion or by a third reviewer (Z. N. Cheng). The following critical data were extracted from each eligible study: first author, publication year, ethnicity, number of patients, median age, variation category, treatment modality, response criteria and genotype data.2 test in this meta-analysis. The heterogeneity across studies was identified by a significant Q test (P<0.10) or by I2>50%; thus, the random effects model was selected for the evaluation of each investigation with combined ORs. In contrast, the fixed effects model was used for P>0.10 from Q test or for I2<50%. Subgroup analysis was undertaken of combinations of anti-angiogenic agents in chemotherapy regimens. To evaluate the stability of the outcomes, sequential exclusion of individual studies was performed in the sensitivity analysis [ORs and 95% CIs were used to evaluate the associations of VEGF gene polymorphisms with response to chemotherapy in CRC, using Review Manager software, version 5.3 (provided by the Cochrane Collaboration), and statistical significance of the OR was ascertained with a P value from the Z-test less than 0.05. Six contrasts for the VEGF -2578 C/A polymorphism were evaluated: comparison of the A allele with C allele; comparison of CC+CA vs. AA; comparison of CC vs. CA+AA; comparison of CC vs. AA; comparison of CC vs. CA; and comparison of CA vs. AA. An evaluation of similar comparison models was also performed in the other three VEGF gene polymorphisms, including VEGF -460 C/T, VEGF +405 G/C and VEGF +936 C/T. Applicability to the effects models depended on the degree of between-study heterogeneity, which was estimated by Cochran's Q test and the Ianalysis . The effanalysis . Two sidA total of 35 relevant studies were retrieved from an initial search of the PubMed, Embase and Medline databases. After removing duplicates and reviews, 18 full-text articles were assessed for eligibility. Finally, 7 studies met the inclusion criteria and were included in the meta-analysis –23. StudThe characteristics of the selected studies are listed in 2 = 75%, P heterogeneity = 0.02), so a random-effects model was used. However, no evidence of heterogeneity was found in the other five comparison models of the C allele vs. A allele , CC vs. CA+AA , CC+CA vs. AA , CC vs. AA , or CC vs. CA , so the fixed-effects model was applied for those genetic models , GG vs. GC+CC , and GG vs. GC , random-effects models were utilized. The remaining genetic models all used fixed-effects models. In the overall analysis, no significant associations were found in any of the comparison models including the G allele vs. the C allele , GG vs. GC+CC , GG+GC vs. CC , GG vs. CC , GG vs. GC , or GC vs. CC   . In addi = 0.41) .A total of 4 studies were included in this meta-analysis. The effects model was finally selected as the fixed-effects model for all six comparison models, mainly because no between-study heterogeneity was found in these genetic models. No significant associations between the VEGF +936 C/T polymorphism and response to chemotherapy in CRC were identified in any comparison models, including C allele vs. T allele , CC vs. CT+TT , CC+CT vs. TT , CC vs. TT , CC vs. CT , or CT vs. TT  . AdditioIndividual studies were consecutively excluded in the sensitivity analysis to investigate whether the obtained results were robust. The analysis showed that the results obtained in the meta-analysis were statistically robust, because the corresponding combined ORs in all of the separate subgroup analyses were relatively stable when deleting any individual study. Publication bias was evaluated with both visual assessment of Begg’s funnel plot and Egger’s test in the meta-analysis. As illustrated in Some published studies have reported inconclusive results about the associations between polymorphisms in the VEGF gene and response to chemotherapy in CRC, probably due to limited predictive ability with relatively small sample sizes. For this reason, a meta-analysis was performed to obtain a comprehensive conclusion on the basis of pooled data from all 7 eligible studies. In this meta-analysis, 4 common SNPs in the VEGF gene were systematically investigated for their associations with response to chemotherapy in CRC. As described in Recently, new biological therapies employing anti-angiogenic agents, including EGFR and VEGF inhibitors, such as cetuximab and bevacizumab, respectively, have been combined with existing chemotherapy regimens because the optimal first-line treatment is no longer chemotherapy alone but a combination with new biological therapies . TherefoMoreover, a significant association with responsiveness to chemotherapy in CRC was identified in the CC vs. CA model of the VEGF -2578 C/A polymorphism and the CC+CT vs. TT model of the VEGF -460 C/T polymorphism. By undertaking subgroup analyses with regard to combinations of anti-angiogenic agents in chemotherapy strategies, we found that associations were only significant in the subgroups excluding anti-angiogenic agents, while negative results were shown in subgroups of including the anti-angiogenic agents Figs  and 4. TAlthough surprising but valuable information was initially obtained in this meta-analysis, this meta-analysis was nevertheless limited due to some deficiencies. First, the limited numbers of both the studies and subjects might have provided insufficient statistical power to evaluate the associations between VEGF polymorphisms and responsiveness to chemotherapy. Second, the heterogeneity of chemotherapeutic regimens might have affected the accuracy of the analysis results. Although the limited number of related studies made it difficult to perform a meta-analysis in the present study when stratified according to chemotherapeutic regimens, a more accurate stratification should be undertaken in the future on the basis of more related studies being published. Third, the sources of inter-study heterogeneity could not be addressed for most of the polymorphisms. Fourth, although there was no evident publication bias identified, potential bias might have distorted the results of the meta-analysis. Finally, relevant effects caused by other environmental factors were difficult to estimate due to publication limitations or incomplete raw data.Although the above limitations existed, this initial meta-analysis of the association between VEGF polymorphisms and responsiveness to chemotherapy in CRC was statistically more persuading than any single study. It concluded that the CC vs. CA model of the VEGF -2578 C/A polymorphism and the CC+CT vs. TT model of the VEGF -460 C/T polymorphism might be predictive factors in responsiveness to chemotherapy in CRC. However, SNPs in the VEGF gene lack sufficient predictive ability as biomarkers to identify whether patients with CRC should add anti-angiogenic agents to their chemotherapy regimes. To assess more accurately the associations between VEGF polymorphisms and responsiveness to chemotherapy in CRC, further studies conducted in standardized and unbiased manner are required.S1 File(DOC)Click here for additional data file.S2 File(DOC)Click here for additional data file.S3 File(DOC)Click here for additional data file.


I2 = 1.0%, P = 0.416). A random-effect analysis comprising seven studies showed high beta-carotene level in serum or plasma was associated with a significant lower risk of all-cause mortality  with low heterogeneity . No evidence of publication bias was detected by Begg’s and Egger’s regression tests. In conclusion, dietary or circulating beta-carotene was inversely associated with risk of all-cause mortality. More studies should be conducted to clarify the dose-response relationship between beta-carotene and all-cause mortality.Observational studies evaluating the relation between dietary or circulating level of beta-carotene and risk of total mortality yielded inconsistent results. We conducted a comprehensive search on publications of PubMed and EMBASE up to 31 March 2016. Random effect models were used to combine the results. Potential publication bias was assessed using Egger’s and Begg’s test. Seven studies that evaluated dietary beta-carotene intake in relation to overall mortality, indicated that a higher intake of beta-carotene was related to a significant lower risk of all-cause mortality  with no evidence of heterogeneity between studies ( It is widely hypothesized that the beta-carotene rich in green leafy vegetables and other orange-colored plant may prevent oxidative damage1In vitro animal studies have demonstrated that beta-carotene may prevent oxidative damage by counteracting the effects of free radicals689Due to inconclusive data on the effect of dietary beta-carotene intake or circulating beta-carotene levels on risk of all-cause mortality in general healthy population, it is critically important to clarify these associations in general population. Therefore, to evaluate these associations and summarize available observational evidence, we attempted to conduct a meta-analysis of prospective studies on this topic.91415161756781019202122232425I2 = 1.0%, P = 0.416) under a random-effect model   .Subgroup analyses were performed across a number of key study characteristics . The sigI2 = 0.0%, P = 0.458).To further confirm the robustness of the results, we conducted sensitivity analyses by excluding sequentially one study at a time. The pooled risks ranged from a low estimate of 0.67(95%CI: 0.59–0.76) when removing the study by Bates10P = 0.764; Egger’s test: P = 0.567). Similarly, no publication bias was observed by the funnel plot, Egger’s regression test (P = 0.209), or by Begg’s rank correlation test (P = 0.368) in the meta-analysis on association between circulating concentration of beta-carotene and risk of all-cause mortality. Funnel plots are provided in In analysis of dietary intake of beta-carotene and all-cause mortality, visual inspection of Begg’s and Egger’s regression tests provided no evidence of publication bias  was a prospective cohort, case-cohort, or nested case-control study conducted in general healthy population, (2) reported dietary intake of beta-carotene or blood beta-carotene concentration as exposure status, (3) presented total mortality as the outcome of interest, (4) provided information about relative risk (RR) and the corresponding 95% confidence interval (CI) or data necessary to calculate these. In addition, we manually reviewed the references list of all previous reviews, relevant meta-analysis and the studies that were included in this analysis to identify other eligible articles that were not found in preliminary document retrieval. When the studies with same population were published repeatedly, priority was given to the publication with largest number of cases or which with most applicable estimates.Two investigators (Long-Gang Zhao and Qing-Li Zhang) independently reviewed all available studies and extracted data with a standard collection form. Any discrepancies were discussed and resolved by the two authors. The following characteristics in the identified studies were recorded: first author’s last name, year of publication, country in which the study was conducted, study name, study period, response rate, follow-up rate, sample size, total number of death, population age at baseline, method used to assess dietary intake of beta-carotene, median or mean dietary intake or circulating level of beta-carotene, categories of dietary intake of beta-carotene or circulating concentration and the RRs or hazards ratio (HRs) and 95%CIs for all-cause mortality associated with those categories, gender of study population, and covariates included for adjustment in multivariable regression models. If multiple estimates were provided, results with multivariable-adjusted risk estimates that were adjusted for most potential confounding factors in original studies were adopted.et al.1719For two studies reported the RR for per standard deviation10χ2) and I2 statistics. Meta-regression was performed to investigate potential sources of heterogeneity between studies. Critical heterogeneity was defined when I2 > 50% or Pfor Q statistic < 0.10In order to examine the association between beta-carotene intake, circulating beta-carotene concentration and risk of all-cause mortality, we calculated pooled RR and 95% CI comparing the highest with the lowest category of beta-carotene intake or blood concentration. We adopted the random-effect model to account for variation between studies proposed by DerSimonian and LairdWe did not assess study quality but we conducted stratified analysis by such specific study characteristics as duration of follow-up, cohort size, exposure measurement types, population age at baseline, median exposure level and adjustment for confounders , which are indicators of study quality. Furthermore, we performed sensitivity analyses by excluding one study at a time to test the influence of individual study on the pooled estimate. Potential publication bias was evaluated by visual inspection of funnel plot and formal testing using Egger’s and Begg’s testsP values were two-sided with a significance level of 0.05 if not specified.All statistical analyses were carried out with Stata (version 13.0). How to cite this article: Zhao, L.-G. et al. Dietary, circulating beta-carotene and risk of all-cause mortality: a meta-analysis from prospective studies. Sci. Rep.6, 26983; doi: 10.1038/srep26983 (2016).


MTHFR) C677T and methionine synthase reductase (MTRR) A66G polymorphisms with being overweight/obesity. However, the results are still controversial. We therefore conducted a case-control study (517 cases and 741 controls) in a Chinese Han population and then performed a meta-analysis by combining previous studies . In our case-control study, the MTHFR C677T polymorphism was not significantly associated with being overweight/obesity when examining homozygous codominant, heterozygous codominant, dominant, recessive and allelic genetic models. The following meta-analysis confirmed our case-control results. Heterogeneity was minimal in the overall analysis, and sensitivity analyses and publication bias tests indicated that the meta-analytic results were reliable. Similarly, both the case-control study and meta-analysis found no significant association between the MTRR A66G polymorphism and being overweight/obesity. However, sensitivity analyses showed that the associations between the MTRR A66G polymorphism and being overweight/obesity became significant in the dominant, heterozygous codominant and allelic models after excluding our case-control study. The results from our case-control study and meta-analysis suggest that both of the two polymorphisms are not associated with being overweight/obesity. Further large-scale population-based studies, especially for the MTRR A66G polymorphism, are still needed to confirm or refute our findings.Several studies have examined the associations of methylenetetrahydrofolate reductase ( Being overweight/obesity has become a serious public health problem worldwide, which could increase the likelihood of many diseases, such as diabetes mellitus, cardiovascular diseases and some cancers . AccordiMTHFR) gene and the A66G polymorphism in the methionine synthase reductase (MTRR) gene were assessed as potential candidates [MTHFR irreversibly catalyzes the conversion of 5,10-methylenetetrahydrofolate to 5-methyltetrahydrofolate, which serves as a methyl donor in the remethylation of homocysteine to methionine [MTRR is responsible for the remethylation of homocysteine to methionine via a vitamin B12-dependent reaction [MTHFR C677T and MTRR A66G polymorphisms can affect the activities of their corresponding enzymes and ultimately lead to elevated homocysteine levels [In the past few decades, many genes and polymorphisms have been hypothesized to be involved in the pathogenesis of being overweight/obesity . Among tndidates ,10,11,12thionine . The MTRreaction . Both the levels ,16,17. Se levels ,19,20. Te levels ,22. Recee levels ,24,25.MTHFR C677T and MTRR A66G polymorphisms with being overweight/obesity, but with conflicting results in different ethnic populations [MTHFR C677T polymorphism with being overweight/obesity was observed in Danish [MTHFR gene with BMI, weight or hip circumference [MTRR A66G polymorphism was not as well studied as the MTHFR C677T polymorphism. We systematically searched seven databases and found that only three studies have evaluated the relationship of the MTRR A66G polymorphism with obesity to date. The MTRR 66G allele was found to be significantly associated with an increased risk of obesity among Italians [MTHFR 677T allele and 677TT genotype was very high in the Chinese population, especially in northerners [MTHFR C677T and MTRR A66G polymorphisms and overweight/obesity risk.Prior studies have explored the relationships of the ulations ,10,11,12n Danish , Iraniann Danish , Saudi [n Danish  and Thain Danish  populatimference . HoweverItalians  and IndiItalians . In Chintherners . Thus, tThe demographic and clinical characteristics of the study subjects are summarized in MTHFR C677T and MTRR A66G polymorphisms are shown in MTHFR C677T polymorphism, the genotype distribution in the control group was consistent with Hardy–Weinberg equilibrium (HWE) (p = 0.662). The 677T allele and 677TT genotype frequencies were not significantly different between cases and controls . Under the multivariate logistic regression model adjusted for age and sex, no significant association was found between the MTHFR C677T polymorphism and being overweight/obesity when examining homozygous codominant (odds ratio (OR) = 1.06, 95% confidence interval (CI) = 0.75–1.51, p = 0.738), heterozygous codominant , dominant , recessive  and allelic  models , which might be attributable to low 66GG genotype frequency and, selection bias in controls owing to a hospital-based case-control study, and there might be a selective pressure acting upon this gene locus. The frequencies of the 66G allele and 66GG genotype were not significantly different between cases and controls . Logistic regression analysis also showed no significant relationship between the polymorphism and being overweight/obesity in homozygous codominant , heterozygous codominant , dominant , recessive  and allelic  models  models . For the) models .MTHFR C677T polymorphism. The pooled results indicated that the MTHFR C677T polymorphism was not significantly associated with the risk of being overweight/obesity in all genetic models  (p > 0.05).Thirteen studies were included in the meta-analysis of the = 0.077) . Between= 0.077) . Sensiti= 0.077) , and staMTRR A66G polymorphism. The pooled results showed that the MTRR A66G polymorphism was not significantly associated with the risk of being overweight/obesity in all of the tested genetic models   . Moderat= 0.055) . Due to MTHFR C677T and MTRR A66G) in homocysteine metabolism genes with overweight/obesity susceptibility.Homocysteine plays a pivotal role in cell metabolism by virtue of its involvement in the transfer of methyl groups in the activated methyl cycle. This cycle is in charge of global and gene promoter-specific DNA methylation, which is one of several epigenetic mechanisms involved in the regulation of gene expression ,22,24,25MTHFR C677T polymorphism was not significantly associated with being overweight/obesity, which corroborates earlier observations in Danish [MTHFR 677T allele carriers had an increased risk of being overweight/obesity. Furthermore, Tabassum and coworkers reported that the 677T allele was associated with a 1.24-fold increased risk of obesity in Indian children [MTRR A66G polymorphism and overweight/obesity susceptibility. In contrast to our observation, Terruzzi and colleagues reported that Italian adults carrying the 66AG genotype had a 2.42-fold higher risk of obesity than 66AA carriers [Initially, we carried out a case-control study in the Han population originating from Tianjin municipality in northern China. To the best of our knowledge, this is the first study exploring the relationships of the two polymorphisms with being overweight/obesity in a Chinese Han population. We observed that the n Danish , Iraniann Danish , Saudi [n Danish  populatin Danish  showed tchildren , and Lamchildren . We alsocarriers . Anothercarriers .MTHFR C677T and MTRR A66G polymorphisms with being overweight/obesity. For the MTHFR C677T polymorphism, thirteen studies with 5431 cases and 24,896 controls were finally included in the meta-analysis. The overall pooled results showed no significant association of the MTHFR C677T polymorphism with overweight/obesity susceptibility and with minimal observed heterogeneity, which confirmed our case-control observations in the Chinese Han population. Subsequent stratified analyses based on ethnicity, source of controls, genotyping method and sample size also did not observe any significant association in all of the subgroups. Sensitivity analyses revealed that these pooled results were reliable, and a publication bias test showed little evidence of publication bias. All of these results indicated that the MTHFR C677T polymorphism might not be a genetic risk factor for being overweight/obesity. For the MTRR A66G polymorphism, only three studies with 1045 cases and 1614 controls were included in the meta-analysis. The overall pooled results also showed no significant association of the MTRR A66G polymorphism with being overweight/obesity. Due to the limited number of studies, we did not perform subgroup analysis, meta-regression analysis and publication bias assessment. However, we performed sensitivity analyses and found that the association between the MTRR A66G polymorphism and being overweight/obesity became significant in three genetic models after excluding our present case-control study. This indicates that our case-control study had a great impact on the overall estimates. It must be noted that the MTRR A66G polymorphism was not in HWE in the controls, which could have affected the results of our case-control study. Thus, a representative sample is still needed to verify a possible association of this polymorphism with being overweight/obesity in the Chinese population, and the meta-analytic results for the MTRR A66G polymorphism should be interpreted with great caution.Meta-analysis is the most commonly used statistical technique in medical research, which combines several individual studies in an effort to achieve higher statistical power, to improve the precision of estimates and to settle uncertainty from conflicting individual results . Given tMTHFR gene polymorphism and reduced muscle mass [MTHFR C677T polymorphism had sufficient statistical power and confirmed the results of our case-control study; however, the meta-analysis on the MTRR A66G polymorphism is underpowered, because of the small sample size. Thirdly, the subjects of our case-control study were from one hospital, which might not possess adequate representation. Fourthly, the included articles were limited to those published in Chinese and English, thus publication bias might have occurred, although funnel plots and Begg’s test showed no evidence of publication bias. Despite these limitations, our study still had its own special advantages. Firstly, it explored the associations of the MTHFR C677T and MTRR A66G polymorphisms with being overweight/obesity for the first time in the Chinese Han population. Secondly, it adopted a comprehensive analysis strategy by integrating case-control study and meta-analysis together, thus enlarging the sample size and strongly enhancing the study power. Thirdly, the heterogeneity test, publication bias assessment and sensitivity analyses indicated that our results for the MTHFR C677T polymorphism were statistically robust and reliable.In interpreting the findings of the current study, several limitations should be acknowledged. Firstly, being overweight/obesity was defined by BMI in our case-control study, which is limited by not being able to distinguish between fat mass and lean or bone mass, and it does not reflect fat distribution. In addition, lack of information on other anthropometric measurements and direct measurements of body composition precluded us from providing a more comprehensive estimation of the two polymorphisms with body composition; especially, several studies have suggested a link between cle mass ,32. SecoBetween October 2008 and February 2011, a total of 517 overweight/obesity subjects and 741 normal weight subjects who took regular health examinations at the physical examination center of Dagang Oil Field General Hospital were recruited into our study. Participants with non-communicable disease history, endocrine diseases and any other diseases that might influence our results were excluded. According to the Guidelines on the Prevention and Management of Overweight and Obesity in Adults: China, normal weight was defined as a BMI <24.0, overweight was defined as 24 ≤ BMI < 28, and obesity was defined as a BMI ≥28. All patients and controls were unrelated and of the same ethnic background. The study was conducted in accordance with the principles stipulated by the Declaration of Helsinki and were approved by the ethics review committee of China Medical University .2). Blood pressure was measured while subjects were in the sitting position after 15 min of rest. The average of three measurements was recorded. At the same time, participants were asked for permission to store a blood sample for biochemical analysis and a buccal cell sample for genetic analysis. The levels of TC, HDL-C, LDL-C, TG and FBG in samples were measured using a Hitachi autoanalyzer .Body weight, height and WC were measured using a standard scale with light clothing and barefoot after an overnight fast. BMI was calculated as weight in kilograms divided by the square of height in meters . Genotyping for the eviously .MTHFR C677T and MTRR A66G allele and genotype frequencies in the cases and controls were calculated by direct counting. Chi-square analysis was performed to identify departures from HWE and to compare the difference between the two groups with respect to allelic and genotypic frequencies. The unconditional logistic regression analysis was performed to estimate the effects of the two polymorphisms on overweight/obesity risk after adjustment for age and gender. OR with 95% CI was calculated to estimate the relative risk of the different genotypes and alleles. A p-value below 0.05 was taken as statistically significant. These analyses were performed using SAS Version 9.2 .The MTHFR C677T and/or MTRR A66G polymorphisms with being overweight and/or obesity. Search strategies were based on combinations of the following key words: obesity, adiposity, obese, overweight, MTHFR, methylenetetrahydrofolate reductase, MTRR, methionine synthase reductase, allele, gene, genotype, variant, variation and polymorphism. The reference lists of retrieved articles were hand-searched to find potentially eligible studies. All of the included studies had to meet the following criteria: (1) evaluating the associations of the MTHFR C677T and/or MTRR A66G polymorphisms with being overweight and/or obesity; (2) case-control or cohort studies in design; and (3) OR with 95% CI could be obtained or calculated with genotype data.We systematically searched three English  and four Chinese databases , Wanfang, China Biological Medicine Database (CBM) and Chongqing VIP Chinese Science and Technology Periodical Database (VIP)) for studies exploring the associations of the Two curators independently extracted the following information from each included study: first author’s name, year of publication, ethnicity and country of study population, source of controls, genotyping method, diagnostic criteria of overweight and obesity, mean age, gender proportion, counts of alleles and genotypes and the numbers of cases and controls. If there was any discrepancy, an agreement was reached by discussion between the investigators.MTHFR C677T and MTRR A66G polymorphisms with being overweight/obesity were assessed by calculating pooled ORs with their corresponding 95% CIs under the homozygous codominant, heterozygous codominant, dominant, recessive and allelic models [I2 statistic [p < 0.05 for the Q-test or I2 > 50% was considered with significant heterogeneity. Once the effects were assumed to be heterogeneous, the random effects model was used; otherwise, the fixed effects model was applied [p-value less than 0.05 was considered to be statistically significant.The HWE in the control groups was tested again. The associations of the c models . The sigtatistic ,34. p <  applied . Subgrou applied . Publica applied . All staMTHFR C677T nor MTRR A66G polymorphism is associated with overweight/obesity risk. However, considering the limitations mentioned before, further large-scale population-based studies, especially for the MTRR A66G polymorphism, are still needed to confirm or refute our findings. We hope that these results will provide background data for the future study of obesity pathogenesis and will contribute to genetic marker screening.In conclusion, our case-control study in combination with the following meta-analysis suggests that neither the


Supplemental Digital Content is available in the text The association between alcohol consumption and Barrett's esophagus (BE) remained uncertain and controversial in the previous studies. We performed a meta-analysis of observational studies to clarify the association.We searched PubMed, Web of Science, and Embase for studies on alcohol consumption and risk of BE published before February 2015. A total of 20 studies reporting the association between alcohol consumption and the risk of BE were identified. Subgroup analyses, meta-regression analyses, sensitivity analyses, and publication bias tests were also performed. Several results from individual studies were pooled using a dose–response meta-analysis.2 = 48.60%) in our study. In subgroup analysis, alcohol consumption was associated with an increased risk of BE in men  and Asian population . In beverage-specific consumption analysis, liquor was associated with an increased risk of BE . Multivariate meta-regression analysis suggested that geographic area, and adjusted age, sex, body mass index, and smoke, might explain 70.75% of the heterogeneity between the studies. We also found the inverse association  between alcohol consumption and BE among subjects when compared with population controls.A total of 20 studies involving 45,181 participants and 4432 patients of BE were included in the meta-analysis. No association was found between alcohol consumption and BE (relative risk [RR] = 1.10, 95% confidence interval [CI] 0.96–1.27, IOverall, there was no significant association between alcohol consumption and BE. Alcohol consumption may be a risk factor of BE in men and Asian population, and liquor consumption may also increase the risk of BE. Significant inverse association was observed between alcohol consumption and BE, for comparisons with population controls. Previous studies have showed a weak association between alcohol drinking and EAC.16 However, recent studies of beverage-specific alcohol consumption also reported lower risk of BE and EAC associated with modest wine drinking,20 whereas others reported higher risk associated with total alcohol9 and liquor consumption.21 It is unclear whether these disparate results are due to measurement error in the assessment of alcohol consumption, or methodological differences in exposure definitions, or differences between the study populations, or effect modification by known causal factors for BE, or other aspects of the study design or analysis.Important risk factors for BE include gastroesophageal reflux disease (GERD) symptoms, abdominal obesity, tobacco use, and male sex.To date, no meta-analysis of the relationship between alcohol drinking and BE has been performed. With the aim to evaluate the effect of alcohol on the risk of BE, we therefore conducted a comprehensive meta-analysis of published case-control and cohort studies.22This review was performed according to the Meta-analysis of Observational Studies in Epidemiology (MOOSE) guidelines.We carried out a literature search using the terms “Barrett's esophagus” or “Barrett's epithelium” or “Barrett syndrome” with “ethanol” or “alcohol” or “alcoholic beverages” to search PubMed, Embase and Web of Science databases for identification of articles published from 1976 to March 31, 2015. We also conducted manual searches of the reference lists of all the relevant original and review articles to identify additional eligible studies. A search for unpublished literature was not performed and authors were not contacted for missing data. Studies were included if they met the following inclusion criteria: studies used a case-control, nested case-control, or cohort study design; BE was diagnosed by the histologic finding of intestinal metaplasia within an endoscopic identified columnar-lined esophagus; and the risk point estimate was reported as relative risk [RR] or odds ratio [OR] and the corresponding 95% confidence intervals (CIs), or sufficient information provided to calculate these estimates. We excluded studies that did not meet the inclusion criteria. Specifically, studies were excluded for the following reasons: studies looked at endoscopic suspected BE patients; studies were present as proceedings and were not published as original articles. The literature search and inclusion or exclusion was independently undertaken by 2 investigators (QX and WG) using a standardized approach. Any inconsistencies between these 2 investigators were settled by the third investigator (XS) until a consensus was reached. Institutional review board approval and patient consent were not required for this meta-analysis of observational studies.23 The NOS uses 2 different tools for case-control and cohort studies, and consists of 3 parameters of quality: selection, comparability, and exposure/outcome assessment. The NOS has developed a “star system”  for the assessment of a maximum of 4 points for selection. A total score of 7 or greater was used to indicate high-quality studies, and a total score of 6 or lower indicated low-quality studies.We performed the data extraction via a standardized data extraction form, collecting information on the author publication year, study location, study design, source of study population, sample size, assessment of alcohol consumption, age of subjects, proportion of males, follow-up time, the number of cases/noncases or person-year data, type of controls, effect estimate and its corresponding 95% CIs, and covariates adjusted in the statistical analysis. Quality assessment of each selected study was conducted by 2 investigators (QX and WG) using the Newcastle–Ottawa Scale (NOS).http://links.lww.com/MD/A372). We used fixed-effect models to evaluate the pooled RR with its 95% CI if there was no evidence of heterogeneity; otherwise, we used random-effect model.25 Next, we conducted a dose–response analysis in order to take into account the correlation between the log of RRs across categories of alcohol consumption for which details of the methods used have been described by Orsini.27 Only studies that reported RRs with their corresponding 95% CIs, for at least 3 quantitative categories were included. We examined a potential nonlinear dose–response relationship between alcohol consumption and BE among those studies reporting level-specific RR estimates with random-effects models. The P value for nonlinearity was calculated by testing the null hypothesis that the coefficient of the second spline was equal to zero. To investigate the sources of heterogeneity between the results of different studies, we carried out the following tests: heterogeneity tests, subgroup analysis, meta-regression analysis, and sensitivity analysis.29 The Cochran Q test and I2 statistic were used to explore the heterogeneity among studies.30 We considered P value was <0.10, and I2 value was >50% significantly statistical heterogeneity.31 Finally, by using the same methodology as for the subgroup analysis, we conducted stratified analyses by categories of sex, beverage type, geographic area, control type, alcohol consumption level, NOS score, adjusted age, adjusted sex, adjusted body mass index (BMI), and adjusted smoke to assess potential effect modification. Univariate meta-regression analysis was conducted first, after which the variables that were significant at the 0.1 level were entered into the multivariable model. To identify potentially influential studies, sensitivity analysis was also performed to examine whether the effect estimate was robust by repeating the random-effects meta-analysis after omitting 1 study at a time.We examined the relationship between alcohol consumption and risk of BE on the basis of the RRs and 95% CIs (estimated by the OR and its 95% CIs in case-control and the hazard ratio and its 95% CIs in cohort studies) reported in each study. We used adjusted risk estimates whenever it is available; otherwise, we utilized or computed the unadjusted RRs. Because different measurement units were used to express alcohol consumption, we converted alcohol consumption levels into grams of ethanol per day for which details are available online . A 2-tailed P value <0.05 was considered statistically significant.Publication bias was assessed by the Egger regression test and Begg test together with the visual inspection of the funnel plot.34 Finally, 20 studies involving 45,181 participants and 4432 patients of BE were included in the meta-analysis after detailed evaluations. Among 20 studies, 12 case-control studies,44 8 cohort studies,50 and 6 studies reporting categories of alcohol consumption were included to conduct the dose–response analysis of the relationship between liquor consumption and the risk of BE.49 Five records from 4 studies were included to conduct the dose–response analysis of the relationship between total alcohol consumption and the risk of BE for comparisons with population-based controls.49 The general characteristics of the included studies are shown in Table The study selection process is shown in Figure P = 0.007, I2 = 48.60%) and no publication bias was found  and 1.31 (95% CI 0.98–1.75) for cohort studies, with heterogeneity , respectively.Figure 2 = 0.00%). The dose–response meta-analysis did not show evidence of a nonlinear relationship between alcohol and risk of BE (P = 0.632). Also, no linear relationship was observed  for every 5 g/d increase in alcohol. We failed to reveal consistent associations between beer, wine, spirits, and the risk of BE. Nevertheless, we found that there was an inverse association  for BE among subjects with GERD when compared with population controls in 6 records from 5 studies,50 which indicated that there might be a U-shaped nonlinear trend between alcohol consumption and risk of BE  and Asian population . We evaluated whether adjusted age, sex, BMI, and smoke modified the association between alcohol consumption and the risk of BE (Table 2 = 25.40%), unadjusted sex , unadjusted BMI , and unadjusted smoke .Alcohol consumption was associated with an increased risk of BE in men , adjusted age (P = 0.027), adjusted sex (P = 0.025), adjusted BMI (P = 0.066), and adjusted smoke (P = 0.026) were significant at the level of 0.1. Thus, the above 5 covariates were entered into the multivariate meta-regression analysis whose results are shown in Table 2 changed from 0.0456 to 0.01334 after including these 5 covariates in the model, which means that 70.75% of heterogeneity between the studies can be explained by these covariates.We used publication year, study design, study quality, total participants, male, geographic area, adjusted age, adjusted sex, adjusted BMI, and adjusted smoke as explanatory covariates. Univariate meta-regression analysis was performed first. Results of the univariate analysis are shown in Table 37 The total result was completely different when we excluded this record .The results of the sensitivity analysis in Table Our meta-analysis identified 20 observational studies through a broad search of manually reviewed databases and rigorous inclusion criteria. Findings from this study showed that total alcohol consumption was not a risk factor for BE. In subgroup analysis, alcohol consumption was associated with an increased risk of BE in men and Asian population. We found that alcohol was a risk factor for BE among subjects with GERD by comparing with GERD controls who lack BE on endoscopy. However, compared with population controls, there was an inverse association between alcohol consumption and BE. In beverage type analysis for total alcohol consumption, liquor was associated with an increased risk of BE. The association between alcohol consumption and BE was also modified by other factors, including age, sex, BMI, and smoke.52 which was confirmed by the present study. Our study also found that the risk of BE increased with increasing alcohol consumption in Asian population, which is in accordance with the results of previous studies in Japan and Korea.45 However, the relationship between alcohol consumption and BE was not found in Westerns.53 This might be due to the different disease pattern of BE between Asians and Westerns because most BE patients in Asia are the short-segment type.54 It is not difficult to find that none of the Asian studies included in our meta-analysis had adjusted estimates. Thus, the results that alcohol consumption was associated with increased risk of BE among Asians are possibly due to some potentially confounding factors, which need to be further explored.Studies have indicated that male sex might increase the risk of BE,37 found that compared with population controls, these lifelong nondrinkers and consumption of <41 drinks/wk of total alcohol consumption throughout the life were less likely to have nondysplastic BE. Thrift's another pooled analysis showed that compared with population-based controls, there was a borderline statistically significant inverse association between any alcohol consumption and the incidence of BE.20 A possible explanation for these somewhat discrepant findings might be that most BE patients drink more alcohol in early life, and then slowly reduce the intake as a result of either their discomfort symptoms or diagnosis.Subgroup analysis indicated that there was a statistically significant inverse association for BE among subjects with GERD when compared with population-based controls. A large population-based case-control study conducted by Thrift et al55 Veugelers36 also reported that increased liquor consumption was a risk factor for both GERD and BE. There are several potential mechanisms through which different alcohol type may be associated with BE. First, liquor drinkers are less likely to consume their alcohol beverage with food. Consumption of alcohol without food may directly damage the lining of the esophagus and increase the esophagitis process, whereas mixed liquor consumption cannot increase the risk.56 Another possibility is that liquor consumption is proxy for some unmeasured unhealthy lifestyle, such as eating fewer fruits and vegetables and having high BMI, which in turn explain the significant risk associations, because many studies have reported that frequency of general alcohol consumption and type of beverage are related to many factors.58The association between liquor consumption and BE was first identified by Ritenbaugh.37 Thrift's study found that there was evidence of an inverse trend for nondysplastic BE, nondysplastic BE patients reported lower intakes than population controls, the possibility seemingly protective effect of lifetime alcohol consumption, as BE patients may refrain from alcohol consumption over time after enduring prolonged reflux discomfort.37 Therefore, whether alcohol consumption increased risk of progression of nondysplastic BE to high-grade dysplasia/adenocarcinoma or not need to be further explored.Sensitivity analysis indicates that the association between alcohol consumption and BE is completely different by exclusion of nondysplastic BE study.Several strengths of the current study should be highlighted. The main strength is that it is the first meta-analysis focusing on the association between alcohol consumption and the incidence of BE. Furthermore, the ascertainment of outcome is based on endoscopy and histological finding in all studies, and the majority of studies included evaluate multiple confounders such as age, sex, BMI, smoke, and so on.There are also several potential limitations to the study. First, limited by the observational design, exclusion of potential confounders from other BE risk factors cannot be ruled out. A meta-analysis is not able to address problems with confounding factors that could be inherent in the original studies. However, in most studies included in this meta-analysis, the investigators had adjusted for major potential confounders, including sex, age, BMI, and smoke. Marked heterogeneity is also observed across these studies which may reflect differences in study design, study population, and adjustment for confounders. Nevertheless, we carried out stratified analysis, meta-regression, and sensitivity analysis to explore this potential bias. Another limitation is the different definition of alcohol consumption among studies, which might result in heterogeneity in our meta-analysis. Some studies used the grams of alcohol to weigh the alcohol consumption, whereas others used drinks of alcohol. We converted all measures into grams alcohol per day using the definitions that reported in the studies.In summary, the results of this study suggested that there is no association between total alcohol consumption and BE risk. However, alcohol consumption was associated with an increased risk of BE in men and Asians. In beverage analysis, liquor consumption was associated with an increased risk of BE either. We found that alcohol was a risk factor for BE in GERD patients. However, when compared with population controls, there was an inverse association. The dose–response meta-analysis suggested that there might be a U-shaped nonlinear trend between alcohol consumption and risk of BE, and an alcohol consumption of <23 g/d might have a potential beneficial effect on BE.


FCGR2A, was associated with increased risk of Kawasaki disease (KD). However, subsequent studies on the role of this SNP were limited and controversial.Recent genome-wide association study found rs1801274, a functional single nucleotide polymorphism (SNP) in IgG receptor gene A case-control study was conducted in a Chinese Han population including 428 KD patients and 493 controls to examine the association between rs1801274 and KD susceptibility. A meta-analysis was performed in combination with the relevant published studies to further clarify such an association.Q = 17.30, p = 0.139). Moreover, sensitivity analysis supported the robustness of this meta-analysis.Our case-control study found that rs1801274 was significantly associated with increased risk of KD in the Chinese Han population, with an odds ratio (OR) of 1.58 (95% CI = 0.96–2.62) for the GA genotype and 1.93 (95% CI = 1.16–3.19) for the AA genotype compared with the GG genotype. The result of meta-analysis further demonstrated that the A allele of rs1801274 was significantly correlated with KD risk under the allelic model  without heterogeneity by fixed-effects model analysis (FCGR2A gene is significantly associated with increased risk of KD.These results further confirm that rs1801274 in the  Kawasaki disease , first reported by a Japanese pediatric doctor Tomisaku Kawasaki FCGR2A (Fc gamma receptor IIa). The single nucleotide polymorphism (SNP) rs1801274 in FCGR2A gene encodes an H131R substitution, which may be related to the development of KD.So far, the etiology of KD remains unclear even clinical features indicates an infectious agent may be involved in the pathogenesis of KD. Evidence from epidemiological studies highly suggested that genetic factors play an important role in this process. First, twins and siblings of affected children have a risk of KD 10 times higher than general population FCGR2A is commonly expressed on immune responsive cells like macrophages, neutrophils, monocytes and dendritic cells FCGR2A have been identified.FCGR2A gene was assumed to be a potential marker of KD risk. However, the results from previous studies were inconsistent FCGR2A gene and KD risk in a Chinese Han population from Zhejiang Province in Southeast China. Additionally, a meta-analysis was performed by combining the data of this study with the results from other relevant, previously published case-control studies, and the data of transmission/disequilibrium test (TDT) to further confirm such an association.Given its genetic function, In this present study, a total of 428 KD patients were recruited from April 2009 to September 2012 in Children's Hospital, Zhejiang University in China. All the patients were unrelated children of Chinese Han ethnic origin. The diagnosis of KD was based on the latest diagnostic criteria released by the Japan Kawasaki Disease Research Committee in 2002 This study was approved by the Ethics Committee of Children's Hospital, Zhejiang University School of Medicine, and Huazhong University of Science and Technology (HUST). Written informed consent forms were obtained from parents/guardians of all the participants.Genomic DNA was extracted from 2 ml peripheral blood sample with the RelaxGene Blood DNA System DP319-02  following the manufacturer's instructions. TaqMan SNP Genotyping Assay was used to determine rs1801274 genotypes using a 7900 HT Fast Real-Time PCR System .2 test between patients and controls. Hardy-Weinberg equilibrium (HWE) was tested by a goodness-of-fit χ2 test for the distribution of genotypic frequencies among controls. After adjustment of gender, unconditional logistic regression analyses were performed to calculate the odds ratios (ORs) and 95% confidence intervals (CIs) for the effect of the SNP on KD and CALs in patients and controls, as well as the patients with and without CALs, respectively. The analysis was conducted under dominant, recessive and additive models respectively to avoid the bias of genetic models. All the statistical analyses were performed with SPSS 11.0 software . Power calculation was carried out using Power version 3.0 with the current sample size and genotype frequency. All p values were two-tailed. P<0.05 was considered statistically significant.The gender distribution and genotypic frequencies were estimated by Pearson χrs1801274, Fc gamma receptor, polymorphism, and Kawasaki disease were used without language restriction. References cited in the retrieved articles, relevant studies or review articles on this topic were also included. The selected studies should meet the following criteria: (1) a case-control or family-based TDT study assessing the association between rs1801274 and KD; (2) containing integrated information about genotype or allele frequency for risk estimates, or original data; (3) genotypes in control groups were fit to Hardy-Weinberg equilibrium (p>0.05). Animal studies, commentaries and case reports were excluded. If subjects were overlapped in several studies, only the one with more complete design or larger sample size was selected.Several databases including EMBASE, PubMed, and ISI Web of Science were searched up to August 28, 2013 to identify the relevant case-control and TDT studies on the association between rs1801274 and KD. The search terms The information extracted from each eligible article included the name of first author, year of publication, study group, ethnicity, method of genotyping, design type and diagnostic criteria for KD.2-based Q statistic test was used to estimate between-study heterogeneity. It was considered statistically significant when p<0.1. In this case, a random-effects model, using DerSimonian and Laird method, was further applied to calculate the pooled OR. Otherwise, a fixed-effects model, using Mantel-Haenszel method, was employed. In order to integrate the results from both case-control and TDT studies, the method described by Kazeem and Farrall http://www.r-project.orgp values were from two-tailed test with a significant level at 0.05 except those for heterogeneity. The meta-analysis was carried out using Catmap software V1.6.Cochran's χ2 = 0.000, p = 0.997). Generally, CALs were detected in 29 patients. The statistical power was 0.846 for the current sample size (493/428), and 0.190 for CAL group (399/29).A total of 428 KD patients and 493 healthy controls were included in this case-control study. Characteristics of all the patients and controls were summarized in p = 0.486) in the controls. As shown in The genotyping of SNP rs1801274 was successful in 99.35% of the study subjects (425/428 patients and 493/493 controls). The genotypic frequency distribution was in agreement with HWE . Therefore, a fixed-effects model was employed to calculate the pooled OR . No significant heterogeneity was detected under the fixed-effects model .Publication bias was examined by funnel plot  and regrIn this present study, the individuals carrying the A allele or AA genotype of rs1801274 were at higher risk of developing KD but were not associated with CAL formation significantly. Subsequent meta-analysis further supported the association between this SNP with increased KD risk. Cumulative and sensitivity analyses confirmed the reliability and robustness of our results, and with little publication bias.FCGR2A's binding affinity to IgG indirectly, resulting in activation of FCGR2A signaling pathway and up-regulation of IgG2-opsonized phagocytosis These findings are biologically plausible for the following reasons. First, the original GWAS on KD conducted by Khor et al FCGR2A on the cells with AA genotype had stronger binding affinity to C-reactive protein (CRP), which showed similar functional activities as IgG2 FCGR2A may affect the pathophysiology of coronary artery indirectly. However, the present case-control study in Chinese Han ethnic children did not confirm whether rs1801274 was associated with CAL. This may be due to the lack of enough CAL cases in our study.Moreover, as the most severe complication of KD, CAL is the leading cause of long-term morbidity and mortality of KD patients. A recent study on endothelial function indicated that the G allele of rs1801274 was associated with endothelium-dependent vasodilatation and nitric oxide activity when endothelial cells were stimulated Several limitations should be kept in mind when interpreting the results of this study. First, the sample size was relatively small, which might affect the power to detect the association between this SNP and CAL. Second, KD is a complex disease associated with both genetic and environmental factors. Local environmental data were not included in this analysis which limited the interpretation of these results and evaluation of the gene-environment interaction.FCGR2A is significantly associated with the increased risk of KD.In conclusion, this present study has provided more evidence to support the concept that the functional polymorphism rs1801274 in Figure S1P<0.001.The forest plot for the cumulative meta-analysis. Fixed-effects pooled OR = 1.35, 95% CI = 1.27–1.44, (TIF)Click here for additional data file.Checklist S1PRISMA Checklist.(DOC)Click here for additional data file.File S1FCGR2A gene in patients with CALs and non-CALs in current study. Table S2, FCGR2A rs1801274 risk allele A in Meta analysis (Asian subgroup).Table S3, Sensitivity analysis of the allelic model.Table S1, Polymorphism of rs1801274 of the (DOCX)Click here for additional data file.