The extraordinary accomplishment of decoding the human genome, together with the technological advancement of the later part of the 20th century, has strong implications for the health care, as it provides the opportunity to transform its delivery health from a disease-focused model to one that is personalized, predictive, proactive, precise, and patient-centred. Personalized health care can be broadly defined as a customization of the medical provision that accommodates individual differences in all stages in the process, from prevention, to diagnosis and treatment, to posttreatment follow-up. Despite the fast moving human genome discoveries in a wide proportion of diseases having large public health impact, however, the promise of personalized healthcare has far lagged behind due the complexity involved . In addiThis special issue includes five contributions spanning from genetic association studies to the awareness of citizens on cervical cancer risk factors and screening programs and physician knowledge on genetic tests for diagnosis of hereditary cancer.RFC-1 ge) and Alzheimer's disease in an Italian case-control studies. Similarly, despite the potential for a biological significance of TNFA −308G>A polymorphism on the risk of obesity, M. Barchitta et al. reported the absence of gene-Mediterranean diet interaction on the risk of overweight/obesity among Italian women. A poor adherence to the Mediterranean diet, however, was associated with educational level (less common among those less educated) and younger ages. Lastly, the paper of S. Boccia et al. investigated the effect of CYP, GST, and SULT polymorphisms and their interaction with smoking on the risk of hepatocellular carcinoma. Results show that CYP2E1∗5B and CYP2E1∗6 polymorphisms have a favorable effect on the development of HCC, while polymorphisms of GSTT1 and SULT1A1 might play role in increasing the susceptibility among smokers.The first set of papers is represented by three genetic association studies on chronic diseases. The paper of F. Coppedè et al. reported on the absence of association between the common c.80G>A polymorphism (rs1051266) of the gene coding for the reduced folate carrier (The paper of C. D. Vito et al. investigated the knowledge of cervical cancer risk factors and the predictors of adherence to cervical cancer screening in relation to mass media campaigns in a large study conducted in Italy. Results show that higher educational level, being not married, and living in urban areas were the main independent characteristics associated with a higher level of knowledge of cervical cancer etiology. During the campaign period women had the Pap test more frequently as a consequence of the mass media campaign, strengthening the evidence of the usefulness of media campaigns via local television to foster cervical screening compliance.Lastly, the paper of N. Panic et al. assessed the knowledge and attitudes on genetic tests for breast and colorectal cancer of young medical residents of postgraduates schools. The study represents a follow-up of a larger study of a sample of Italian physicians . ResultsStefania BocciaPaolo BoffettaPaolo Villari


We performed a meta-analysis of cholinesterase inhibitors for patients with Lewy body disorders, such as Parkinson’s disease, Parkinson’s disease dementia, and dementia with Lewy bodies.The meta-analysis included only randomized controlled trials of cholinesterase inhibitors for Lewy body disorders.Seventeen studies (n = 1798) were assessed. Cholinesterase inhibitors significantly improved cognitive function , activities of daily living (SMD = −0.28), and global function (SMD = −0.52) compared with control treatments. Changes in motor function were not significantly different from control treatments. Furthermore, the cholinesterase inhibitor group had a higher all-cause discontinuation , discontinuation due to adverse events , at least one adverse event , nausea , and tremor .Cholinesterase inhibitors appear beneficial for the treatment of Lewy body disorders without detrimental effects on motor function. However, a careful monitoring of treatment compliance and side effects is required. Lewy body disorders, such as Parkinson’s disease (PD), PD dementia (PDD), and dementia with Lewy bodies (DLB), are neurodegenerative diseases characterized by accumulation of Lewy bodies in brain cells . CognitiP < .00001]. P = .003, 3 RCTs, n = 440; 10mg donepezil: WMD = −1.31, 95% CI = −2.53 to −0.09, P = .04, 4 RCTs, n = 450; and 12-mg rivastigmine: WMD = −1.04, 95% CI = −1.65 to −0.43, P = .0009, 2 RCTs, n = 621].A recent meta-analysis showed that cholinesterase inhibitors (ChEIs)  were superior to placebo in improving cognitive function in patients with AD . In addiAs PD is a Lewy body disorder, we performed a meta-analysis of ChEI safety and efficacy for treating patients with Lewy body disorders, including DLB, PDD, CIPD, and PD. This analysis pooled the results of 17 RCTs (involving 1798 patients) using the same methodology as that used in our previous meta-analysis .This meta-analysis was performed according to Preferred Reporting Items for Systematic Reviews and Meta-Analysis guidelines . We systWe included only RCTs of ChEIs for patients with Lewy body disorders. Open-label, nonplacebo-controlled , and crossover studies were included for increasing the sample size. To identify relevant studies, we searched PubMed, Cochrane Library databases, EMBASE, CINAHL, and PsycINFO citations. There were no language restrictions, and we considered all studies published up to July 14, 2015. We used the following key words: “cholinesterase inhibitor,” “donepezil,” “galantamine,” “rivastigmine,” “Lewy,” “Parkinson disease,” or “Parkinson’s disease.” Additional eligible studies were sought by searching the reference lists of the primary articles and relevant reviews.http://ims.cochrane.org/revman). Discrepancies in different coding forms were resolved by discussions between authors (S.M. and T.K.)Two authors (S.M. and T.K.) scrutinized the patient inclusion and exclusion criteria for the identified studies. When data required for the meta-analysis were missing, the first and/or corresponding authors were contacted for additional information, including endpoint scores. Three authors  independently extracted, assessed, and entered the data into Review Manager  or modified ITT data . However, we analyzed the complete set of data for ensuring that the maximum possible information was included .I2 statistic, with values of ≥50% regarded as considerable heterogeneity , placebo-controlled or non-placebo–controlled trial, diagnosis , trial duration , type of ChEI , sample size (>28 per group or ≤28 as 28 was the median number of participants per group), sponsorship (industry sponsored or not), and quality of the trial design . Funnel plots were visually inspected for assessing the possibility of publication bias. We also assessed the methodological qualities of the articles included on the basis of the Cochrane risk of bias criteria .For combining studies, we used the random effects model of ogeneity . In caseThe search yielded a total of 946 references, of which 550 were duplicates . We exclThe 17 RCTs included 1798 patients . The stuP<.0001, I2=68%, 16 comparisons, n=1889) . In contrast, there was no significant difference in cognitive function scores between galantamine and control groups .Pooled ChEIs improved cognitive function scores compared with control treatments  . Visual  n=1889) . For indI2=68%) ; test for subgroup differences, I2=95.6%, P<.00001]. When divided into placebo-controlled and non-placebo–controlled trial subgroups, the same results were found. When divided according to the tested ChEIs, the significant heterogeneity disappeared in donepezil and rivastigmine subgroups ; however, heterogeneity remained in the galantamine subgroup (I2=94%). When divided into DLB, PDD + CIPD, and PD subgroups, there was significant heterogeneity among all subgroups . ChEIs significantly affected cognitive function scores for DLB and PDD+CIPD subgroups . In contrast, there was no effect of ChEIs on cognitive function scores in the PD subgroup . When divided into long duration (>12 weeks) and short duration (≤12 weeks) subgroups, there was significant heterogeneity in both subgroups . When divided into high-quality trial  and “other” subgroups, the significant heterogeneity disappeared in the high-quality trials subgroup (I2=45%) but remained in the “other” subgroup (I2=85). Further, when divided into large and small sample size subgroups, the significant heterogeneity disappeared in the small sample size subgroup (I2=7%) but remained in the large sample size subgroup (I2=75%). When divided into industry and nonindustry subgroups, the significant heterogeneity disappeared in the industry subgroup (I2=48%) but remained in the nonindustry subgroup (I2=89%).There was significant heterogeneity in cognitive function scores among the studies (I2=68%) . TherefoI2=68%) . When dine study ], I2=noP=.03, I2=81%, 13 comparisons, n=1832) . There was also a trend toward improved behavioral disturbance scores for donepezil compared with that for placebo , while galantamine had no significant effect compared with control treatment .Pooled ChEIs improved behavioral disturbance scores compared with control treatments  . Visual  n=1832) . For indI2=81%) ; however, heterogeneity remained in the galantamine subgroup (I2=98%). When divided into DLB, PDD+CIPD, and PD subgroups, there was significant heterogeneity in the DLB and PDD+CIPD subgroups . ChEIs significantly affected behavioral disturbance scores for PDD+CIPD subgroups . In contrast, there was no effect of ChEIs on behavioral disturbance scores in the DLB subgroup . We did not perform meta-analysis on behavioral disturbance scores for PD subgroup, because there was only one relevant study . When divided into long duration and short duration subgroups, there was significant heterogeneity in both subgroups . ChEIs significantly affected behavioral disturbance scores for long duration subgroups . In contrast, there was no effect of ChEIs on behavioral disturbance scores in the short duration subgroups . When divided into high-quality trial and other subgroups, the significant heterogeneity disappeared in the high-quality trials subgroup (I2=21%) but remained in the “other” subgroup (I2=98). ChEIs significantly affected behavioral disturbance scores for high-quality trial subgroups . In contrast, there was no effect of ChEIs on behavioral disturbance scores in the other subgroups . Further, when divided into large and small sample size subgroups, the significant heterogeneity disappeared in the small sample size subgroup (I2=0%) but remained in the large sample size subgroup (I2=84%). ChEIs significantly affected behavioral disturbance scores for large sample size subgroup . In contrast, there was no effect of ChEIs on behavioral disturbance scores in the small sample size subgroup . When divided into industry and nonindustry subgroups, the significant heterogeneity disappeared in the industry subgroup (I2=28%) but remained in the nonindustry subgroup (I2=96%). ChEIs significantly affected behavioral disturbance scores for industry subgroup . In contrast, there was no effect of ChEIs on behavioral disturbance scores in the nonindustry subgroup .There was significant heterogeneity in behavioral disturbance scores among the studies (I2=81%) . TherefoI2=81%) . When dine study ), I2=NAnt study . This stP=.76, I2=8%, 15 comparisons, n=1312)  . Visual  n=1312) . For indPooled ChEIs improved ADL scores compared with placebo (SMD=−0.28) . For indPooled ChEIs improved global function assessment scores compared with placebo (SMD=−0.52) . For indThere was a significantly higher all-cause discontinuation rate in the pooled ChEIs group compared with controls  . RivastiThere was a significantly higher discontinuation rate because of adverse events in the pooled ChEIs group compared with the control group  .There was a significantly higher rate of at least one adverse event between pooled ChEIs and placebo groups  . For indWith respect to individual adverse events, the pooled ChEIs group treatment was associated with a lower incidence of hallucination than placebo group  . The pooTo our knowledge, this is the first comprehensive meta-analysis of RCTs assessing the efficacy and safety of ChEIs for Lewy body disorders. The main results indicate that ChEIs improve cognitive function, behavioral disturbances, ADL, and global function compared with placebo. Moreover, pooled ChEIs did not worsen motor function. According to the effect sizes of individual ChEIs for cognitive function, donepezil was the most effective (SMD=−0.51), followed by rivastigmine (SMD=−0.29), while galantamine had no significant effect compared with placebo. Further, only rivastigmine significantly improved behavioral disturbances compared with placebo; however, even these effects were small (SMD=−0.21). We suggest that the significant heterogeneity among studies of cognitive function and behavioral disturbances was because of variation in quality of the trial design, because the heterogeneity was reduced after excluding one open-label, non-placebo–controlled trial . SensitiWhile these drugs were effective against several cardinal deficits associated with Lewy body diseases, there were also significantly higher rates of all-cause discontinuation, discontinuation because of adverse events, and incidence of at least one adverse event in the pooled ChEIs group compared with controls. Moreover, ChEI treatment was associated with a higher incidence of nausea and tremor compared with placebo. For individual ChEIs, rivastigmine was associated with a higher incidence of vomiting and dizziness compared with placebo.I2=39.6%, P=.20). The second limitation is that characteristics the studies included in the meta-analysis (These conclusions must be considered considering several limitations. The first limitation is that our meta-analysis includes “gray” studies supported by pharmaceutical companies. However, these represented the majority of the retrieved articles. Nonetheless, they appear in peer-reviewed journals. Moreover, there were no significant subgroup differences between industry-sponsored and non-industry–sponsored studies (analysis . The thianalysis . The fouanalysis , reducinIn conclusion, ChEIs are beneficial for the treatment of Lewy body disorders as assessed by multiple scales evaluating cognition, behavioral disturbances, ADL, and global function. Moreover, ChEIs do not worsen motor function. However, a careful monitoring of treatment compliance and side effects is required.Dr. Matsunaga has received speaker’s honoraria from Eisai, Janssen, Novartis, Daiichi Sankyo, Ono, Eli Lilly, Takeda, and Otsuka. Dr. Kishi has received speaker’s honoraria from Abbott, Astellas, Daiichi Sankyo, Dainippon Sumitomo, Eisai, Eli Lilly, GlaxoSmithKline, Yoshitomi, Otsuka, Meiji, Shionogi, Janssen, Novartis, Tanabe-Mitsubishi, and Pfizer. Dr. Yasue has received speaker’s honoraria from Otsuka, Eli Lilly, Janssen, Novartis, and Eisai. Dr. Iwata has received speaker’s honoraria from Astellas, Dainippon Sumitomo, Eli Lilly, GlaxoSmithKline, Janssen, Yoshitomi, Otsuka, Meiji, Shionogi, Novartis, and Pfizer.All authors declare that they have no direct conflicts of interest relevant to this study. No grants or other funding sources were used for this study.


OGG1 gene encodes the enzyme 8-oxoguanine DNA glycosylase which removes oxidatively damaged bases of DNA. Several studies report that the OGG1 Ser326Cys polymorphism increases the risk for cancers of the upper aero-digestive and gastrointestinal tract. However, other studies provide evidence that such an association does not exist. A meta-analysis to assess the role of OGG1 Ser326Cys polymorphism in the cancers of the upper aero-digestive and gastrointestinal tract was therefore undertaken in order to resolve this ambiguity. Seventeen studies were recruited for this meta-analysis after screening 58 articles with a total of 5533 cases and 6834 controls for which the odds ratio with 95 % confidence interval was calculated. Begg’s funnel test and Egger’s test were performed for calculating publication bias. Our study reveals an association between OGG1 Ser326Cys polymorphism and cancer susceptibility of the upper aero-digestive and gastrointestinal tract . Subgroup analysis based on cancer types and ethnicity also revealed the association of OGG1 Ser326Cys polymorphism to the risk for upper aero-digestive and gastrointestinal tract cancers among both the Asian and the Caucasian populations. No risk was however observed for smoking habits and OGG1 Ser326Cys polymorphism. In conclusion, OGG1 Ser326Cys polymorphism may be associated with the increased risk for aero-digestive tract and gastro-intestinal cancers in both Asian and Caucasian populations.Cancers of the upper aero-digestive and gastrointestinal tract are one of the major causes of mortality around the world. DNA repair genes play a vital role in preventing carcinogenesis by maintaining genomic integrity. Polymorphisms in the nucleotide sequence of DNA repair genes are often reported to be associated with an increased risk for different cancers. The  Cancer is a multifarious disease characterised by abrupt growth of cells resulting in abnormal regulation of cell-cycle progression and division Sawyers . The uppOGG1 gene localized on chromosome 3p25 has both DNA glycosylase and apurinic or apyrimidinic (AP) lyase activities. It removes the 8-oxoG lesion by slicing the glycosydic bond between the modified base and the sugar moiety, leaving an abasic apurinic/apyrimidinic (AP) site in DNA  with cysteine (Cys) (Ser326Cys) resulting from a C to G transversion at position 1245 in exon 7 of the OGG1 gene  and a 50 % increase in vegetable and fruit intake are at 50 % decreased risk of developing lung cancer. However, such a decrease in risk was not observed for the other genotypes  articles written in English (4) studies in which full information about genotype distributions are reported (5) Studies in which genotype distribution of control populations are in accordance with Hardy–Weinberg Equilibrium (P > 0.05) (6) only original research articles excluding reviews, letters and case reports. Studies prior to 2007 were excluded because they were mostly constrained to small study size.Articles for the meta-analysis were selected if they met the following criteria: (1) Studies not prior to 2007 (2) case–control study related to the risk of All the data were extracted by two investigators (Das S and Nath S) carefully and independently to maintain accuracy of the data, based on the inclusion criteria above. The information collected from each study are: author’s name, year of publication, number of cases and controls, information of the genotypes of the cases and controls, ethnicity of the study population. The study populations were divided into Asians and Caucasians based on their ethnicity. Based on the location of cancer we have defined two cancer groups as UADT cancer  and GI cancer . In order to study the gene environment interaction, sub group analysis was performed with smoking habits between studies. The analysis was conducted on the basis of non-smokers versus smokers in the cancer patients.Chi square test for Hardy–Weinberg equilibrium was performed in control populations. A P value greater than 0.05 was considered to be in accordance with Hardy–Weinberg equilibrium. Unconditional logistic regression was used to determine the odds ratio (OR) with 95 % confidence interval for all the studies. Cochran’s Q statistic Cochran  was usedOGG1 Ser326Cys polymorphism and the risk of head and neck, oral, pancreatic, gallbladder, colorectal and gastric cancers. Among all the articles identified, only 31 were subjected to further examination out of which only 17   than among the controls (0.35 and 0.05) for the Caucasian population. A similar trend was also observed in the Asian population where the frequency of Ser/Cys and Cys/Cys was higher among the cases (0.43 and 0.15) than the controls (0.41 and 0.12). Results suggest that the polymorphism was a risk for cancer among both the Asian  and the Caucasian  populations for all the three models. Stratification on the basis of cancer types reveals that there was an elevated risk for individuals with this polymorphism for both the GI  and UADT  cancer .Data pertaining to the habit of smoking was available only for a few of the included studies and the genotypes of We observed that the cancers categorized in this study as GI and UADT cancers were of different types, viz. head and neck cancer, oral cancer, nasopharyngeal cancer, esophageal cancer, gastric cancer, colorectal cancer, pancreas cancer and gall bladder cancer Table . Oral caOGG1 Ser326Cys polymorphism  and subsequently the random effect model was selected. We then tried to assess the source of heterogeneity based on ethnicity (Asian and Caucasian), sample size (<500 subjects and ≥500 subjects) and cancer types (GI cancer and UADT cancer). From the analysis it was found that subsequent heterogeneity was contributed by Caucasian population (I2 = 74.3 %) and sample size of below 500 (I2 = 72.6 %) among the studies. While in case of cancer types heterogeneity was found to be contributed by both the GI (I2 = 70.3 %) and UADT (I2 = 64.9 %) cancers.In this study significant heterogeneity was observed for all the models of Begg’s funnel plot Fig.  and EggeOGG1gene encodes the enzyme required for removal of the 8-oxoG adduct from DNA. Although many polymorphisms have been reported in OGG1, the C/G polymorphism at the 326 codon of exon 7 which results in an amino acid substitution from serine to cysteine is of great importance as it has been linked to the increased risk for different cancers. In few studies, the OGG1 protein encoded by the 326Ser variant is found to have more DNA repair activity than that coded by the 326Cys variant indicating the role of this polymorphism in carcinogenesis  we observe a significant risk, clearly indicating the role of the mutant G allele in increasing the risk for UADT and GI cancers.DNA repair genes play an important role in maintaining the genomic integrity of a cell. Polymorphisms in these genes may alter the protein function and thus hamper the DNA repair capacity of an individual (Goode et al. n et al. , Zhang en et al.  and Wangn et al.  did not u et al.  reportedg et al. . Thus, tOGG1 polymorphism increases the risk for cancer in both Asian and Caucasian populations. The maximum risk for Asian population was observed for GG vs CC genotype with a 1.56 fold increase in risk. Similar trend was also observed for the Caucasian population with a 1.4 fold increase in risk. We also observed a significant increase in risk by 1.4 fold among Asians with GG vs CG + CC genotype. The 17 articles included in this study were found to have been performed on either Asian or Caucasian populations. Hence, this meta-analysis reports findings pertaining to only these two ethnicities, although the ethnicity was not a pre-determined criterion for selection.Stratified analysis based on ethnicity reveals that the OGG1 Ser326Cys polymorphism. This may be due to a decrease in sample size resulting from stratification of data based on smoking habit. Our study based on the risk for different types of cancers and OGG1 polymorphism showed highly significant risk for both the head and neck and colorectal cancers which was in accordance with Su et al. (Environmental factors such as smoking in association with different polymorphisms were found to play an important role in the onset of various cancers (Das et al. u et al.  and Wangu et al.  respectiOGG1 on the risk for UADT and GI cancers. Significant heterogeneity was also contributed by sample size and cancer types which indicate that the study recruited for the meta-analysis might have contributed to the increased heterogeneity for which the random effect model was used for calculating the pooled OR. Besides that we have also calculated publication bias by Begg’s funnel plot and Egger’s test, the results of which indicated no such bias in our study. This authenticates that our meta-analysis results are relatively stable.Heterogeneity testing revealed significant heterogeneity among all the three models in our study. However, while trying to assess the source of heterogeneity, we found that maximum heterogeneity was contributed by Caucasian population, indicating a presence of ethnicity specific effect of Although we have conducted our meta-analysis with a very large number of cases and controls the study has some limitations. Firstly, most of the study groups were of Caucasian origin and only few studies pertaining to the Asian population could be recruited for the study. Secondly, we were not able to conduct our study for the other ethnicities, which should be researched in future studies; and thirdly, only a few studies recruited for the meta-analysis contained detailed information on the smoking habits of cancer patients and controls, as a result of which the analysis based on smoking habit must be further validated on a larger study group.OGG1 Ser326Cys polymorphism may be associated with an increased risk for UADT and GI cancers in both Asian and Caucasian populations.In conclusion, our meta-analysis provides evidence that the


Ebola virus disease has reemerged as a major public health crisis in Africa, with isolated cases also observed globally, during the current outbreak.0, which is a measure of the severity of the outbreak, we developed a SEIR (susceptible-exposed-infected-recovered) type deterministic model, and used data from the Centers for Disease Control and Prevention (CDC), for the Ebola outbreak in Liberia and Sierra Leone. Two different data sets are available: one with raw reported data and one with corrected data (as the CDC suspects under-reporting).To estimate the basic reproductive ratio R0 for Liberia resulted to be 1.757 and 1.9 for corrected and uncorrected case data, respectively. For Sierra Leone, R0 resulted to be 1.492 and 1.362 for corrected and uncorrected case data, respectively. In each of the two cases we considered, the estimate for the basic reproductive ratio was initially greater than unity leading to an epidemic outbreak.Using a deterministic ordinary differential equation transmission model for Ebola epidemic, the basic reproductive ratio R0 associated with the 2014 Ebola outbreak, and showed that there is close agreement between our estimates of R0. Analysis of our model also showed that effective isolation is required, with the contact rate in isolation less than one quarter of that for the infected non-isolated population, and that the fraction of high-risk individuals must be brought to less than 10% of the overall susceptible population, in order to bring the value of R0 to less than 1, and hence control the outbreak.We obtained robust estimates for the value of RThe online version of this article (doi:10.1186/s40249-015-0043-3) contains supplementary material, which is available to authorized users. Please see Additional file Ebola virus disease (EVD), named after the Ebola River in Zaire, is known to be a highly contagious disease with a high mortality rate ,2. PreviFiloviridae, genus Ebolavirus. There are five identified subspecies of the Ebolavirus. Four of the five subspecies: i) Ebolavirus (Zaire ebolavirus), EBOV ii) Sudan virus (Sudan ebolavirus), SUDV, iii) Bundibugyo virus (Bundibugyo ebolavirus), BDBV and iv) Ta Forest virus (Ta Forest ebolavirus), TAFV, have caused disease in humans. The fifth, called the Reston virus (Reston ebolavirus), has caused disease in nonhuman primates 0, we useSL(0), SH(0), E(0), I(0) H(0),R(0)) as model parameters, along with the effective contact rate and estimate values for all parameters. Such a technique, however, produces slightly unreliable results. This is explained by the fact that the available epidemic data is restricted to the reported cumulative case number, while the optimization scheme that we employ produces estimates for six variables.Furthermore, we require knowledge of the initial conditions to be used for simulation of the trajectories of Model 1. It is possible to consider the initial conditions (SH(0) = 20000, SL(0) = one million, E(0) = 15, I(0) = 10, H(0) = 0, and R(0) = 0. This initial data is based on the fact that the total population in the region that was under threat of Ebola was one million. Among this population, 2% are at high-risk. Initially there were only a few exposed and infected individuals, and no hospitalized or recovered individuals.There are thus too many degrees of freedom and the ‘best-fit’ may result in unrealistic estimates for the initial conditions used. We use the number of cases first recorded as the initial conditions and restrict ourselves to optimizing only the effective contact rate. The following initial conditions were used; β by minimizing the difference between predictions of Model 1 and the epidemic data. This is implemented by using the fminsearch function in the built-in optimization toolbox in MATLAB. The function fminsearch searches the minimum of a function of several variables. An initial guess of the variables is provided. Then, by using that guess, it searches for the values that are local minimizers of the function. This allows us to estimate the parameter β in order to calculate R0.In the following section we employ ordinary least squares (OLS) estimation to estimate the parameter β is 0.371 and R0 is 1.757. Figure β is 0.3906 and R0 is 1.9.As described in previous section, we fitted the two different data sets, with and without correction for under reporting, for Liberia and Sierra Leone. Figure β is 0.361 and R0 is 1.492, as shown in Figure β becomes 0.344 and R0 is 1.3682, as shown in Figure For Sierra Leone, 0 with the relative risk of the highly susceptible population group and its percentage in the total susceptible population group. We note that less than 10% of the susceptible population should be in the high-risk group in order to bring R0 to less than 1. This is shown in Figure Finally, we study the variation in R0. As shown in Figure 0 to a value less than 1.We also use our model to study the effects of isolation on RWe have developed a model for the transmission dynamics of EVD, incorporating the important factor of the individuals who are at a much greater risk of contracting the disease than the general population in the affected areas. These include frontline health-care workers, family members of EVD patients, and individuals involved in the burial process of deceased EVD patients.0 in this country is 2.53, which is significantly greater.Using data from the WHO and CDC, we have calculated estimates for Liberia and Sierra Leone for the ongoing EVD outbreak. The values are significantly above 1, indicating the severity of the disease. The estimated values for Liberia are consistent with published estimates for the current outbreak ,20,24, w0 to reduce to less than 1, the transmission rate of the isolated individuals should be less than one quarter of that for the non-isolated. This means that strict protocols should be followed at treatment facilities. Further analysis of the model also leads to the conclusion that the fraction of high-risk individuals has to be controlled and must be brought to less than 10% of the overall susceptible population in order to bring R0 to less than 1 and hence control the outbreak.We have also looked at the effect of interventions to control the outbreak. In absence of any vaccine or medication for EVD, the only control measure available is isolation. Ideally during isolation strict control should be in place so that the isolated individuals do not transmit the disease. However, in practice, there is a non-zero risk of transmission from isolated individuals. Our analysis suggests that in order for ROur model is an attempt to capture the most important features of the transmission dynamics of EVD. As an extension of this work, optimal, time-dependent strategies should be developed and advised to public health authorities in order to control the disease.


Cleft lip and palate (CL/P) is one of the most common malformations in humans. Transforming growth factor alpha (TGFA) is a well characterized mammalian growth factor which might contribute to the development of CL/P. This meta-analysis aimed to summarize the association between the TGFA Taq I polymorphisms and CL/P.We retrieved the relevant articles from PubMed, EMBASE, ISI Web of Science and SCOPUS databases. Studies were selected using specific inclusion and exclusion criteria. The odds ratios (ORs) and their 95% confidence intervals (95% CIs) were calculated to assess the association between TGFA Taq I polymorphism and CL/P risk. Meta-analyses were performed on the total data set and separately for the major ethnic groups, disease type and source of control. All analyses were performed using the Stata software.Twenty articles were included in the present analysis. There is a significant association between the TGFA Taq I polymorphism and CL/P . Stratified analyses suggested that the TGFA Taq I polymorphism was significantly associated with CL/P in Caucasians .TGFA Taq I polymorphism may be associated with the risk of CL/P. Facial clefting is one of the most common malformations in humans. Significant differences between populations in the prevalence of cleft lip or palate (CL/P) have been reported, with higher rates found in Asians and American Indians than those observed in Caucasians and Africans. Palate formation is complex, and there are numerous potential untoward possibilities, the most common being delayed shelf horizontalization and inadequate shelf growth .Epidemiologic studies suggest that a number of environmental factors have been examined as risk factors for CL/P, including maternal smoking, exposure to antiepileptic drugs, antiemetic agents and vitamin use during the periconceptual period, maternal metabolic factors, alcohol consumption and exposure to agricultural chemicals . SeveralTaq1 marker in the transforming growth factor alpha (TGFA) gene showing the strongest association. TGFA encoded by a gene mapped at 2p13, is a secretion protein that binds to the epidermal growth factor receptor (EGFR) and is situated at the palate epithelium during palate closing [in vitro and in vivo. Moreover, the continued presence of EGF inhibits the fusion process; TGFA is likely to have similar effects. These biological studies suggest that mutations in the TGFA gene might contribute to the development of CL/P, especially for those mutations that affect the timing of the tissue-specific expression of this gene.The epidemiologic characteristics and risk factors of CL/P are not clear. There is also a strong genetic component to oral clefts. The host susceptibility factors may play an important role in the development of CL/P. Ardinger and colleagues performed the first study to use a case–control design to test candidate genes . They fo closing . EGF/TGFThe TGFA gene shows a restriction fragment length polymorphism when treated with Taq I restriction enzyme. The mutant allele shows a four-base (TAAT) deletion. In this case, it shows a 178-base pair (bp) C1 allele and a 174-bp C2 allele . TGFA TaThe above inconsistent conclusions in the findings of the studies may be attributed to the size of the samples, the ethnic of the sample population and other reasons. In order to contribute to a better understanding of the role of this gene in the occurrence of cleft lip, cleft lip, or cleft lip and palate, we perform an updated meta-analysis on all available case–control studies, combining data together to reach a larger sample size, to get more statistical power to evaluate the association between CL/P susceptibility and TGFA Taq I polymorphism. Understanding the genetic background and etiology of CL/P is essential for both the risk assessment and findings of effective methods for prevention and treatment.We retrieved the articles using the following terms “Transforming growth factor alpha or TGFA” and “cleft lip or cleft palate or cleft lip and palate” from PubMed, Embase, ISI Web of Science and SCOPUS (Last search was updated on October 2013). There was no any language restriction and the age of participants was not considered as selection criteria. We evaluated potentially relevant publications by examining their titles and abstracts and all studies matching the eligible criteria were retrieved.Eligible studies were selected according to the following explicit inclusion criteria: (a) evaluation of the TGFA Taq I polymorphism and CL/P risks, (b) using the methodology of a case–control study to keep the homogeneity between included studies in the meta-analysis.http://www.prisma-statement.org/) was shown in Figure Duplicate and obviously unrelated articles were eliminated by a single author (C.F.). Abstracts of the remaining articles were examined independently by two authors (C.F. and E.Z.) to determine whether the full-text article should be sought. When there were disagreements between CF and EZ in selecting papers, the third author (L.L.) would assess the articles and make the final decision with CF and EZ. A four-phase flow diagram according to Systematic Reviews (The odds ratios (ORs) and their 95% confidence intervals (95% CIs) were calculated to assess the association between TGFA Taq I polymorphism and CL/P. Pooled ORs were obtained from combination of single study by heterozygote comparison (C1C2 vs. C1C1), homozygote comparison (C2C2 vs C1C1), dominant model (C1C2 + C2C2 vs C1C1), recessive model (C2C2 vs C1C2 + C1C1) and allelic model (C2 vs C1), respectively. These comparisons were used to provide more information about the relationship between the polymorphism and the disease, assess the association in different point of view and validate the association by many ways, as well as offer the data for further study on the gene expression. Meta-analyses were performed on the total data set and separately for the source of control and disease type. We investigated the between-study heterogeneity by the Cochran’s Q test and quantified by I2. To obtain summary statistics for ORs of the polymorphism and disease, we performed initial analyses with a fixed-effect model and confirmatory analyses with a random-effect model if there was significant heterogeneity. If there is no heterogeneity, the fixed and random effects models produce similar results, and, if not, the random-effect model usually produces wider CIs than the fixed-effect model. If the P value is >0.05 of the Q test, the summary OR estimate of each study was calculated by the fixed-effect model. Otherwise, the random-effect model was used.t-test as suggested by Egger’s test. All of P values were two-sided and all analyses were performed using the Stata software version 11.0 .We assessed potential publication bias by examining funnel plots and using Egger’s test ,37. A fuDetailed characteristics of each study are summarized in Table NOS results suggested that all of the included studies are high level quality with the score >6.2 = 55.8%; C2C2 + C1C2 vs C1C1: OR = 1.52, 95% CI = 1.15-2.01, P < 0.001 for heterogeneity, I2 = 64.7%; C2 vs C1: OR = 1.41, 95% CI = 1.12-1.78, P < 0.001 for heterogeneity, I2 = 65.2%), but not in the homozygote and recessive model . Meta-analysis results of the association between TGFA Taq I polymorphism and CL/P risk under the heterozygote comparison model (C1C2 versus C1C1), the dominant model (C1C2 + C2C2 versus C1C1), and the allelic model (C2 versus C1) were also shown in Figure A summary of the meta-analysis findings of the association between TGFA Taq I and CL/P risk is provided in Table Stratified analyses were conducted by ethnicity, source of control and disease type. Pooled ORs and 95% CIs of stratified meta-analysis are showed in Table In the subgroup analysis according to disease type, the ORs of the heterozygote comparison, dominant and allelic model with CL/P are statistically significant . For the CP, the significant results were observed in heterozygote comparison, dominant and allelic model .In the subgroup analysis by control characteristics, the ORs of the heterozygote comparison, homozygote, dominant, recessive and allelic model for the hospital-based control are statistically significant . While, no statistically significant association was found in population-based controls.Heterogeneity between studies was observed in overall comparisons and also subgroup analyses. Thus, meta-analyses were performed using random-effect models Table .To explore potential sources of heterogeneity in this meta-analysis, meta-regression analyses were implemented. The covariates included ethnicity and source of control. In all of the heterozygote comparison, homozygote, dominant, recessive and allelic models, above potential factors were probably not the major sources of heterogeneity . The heterogeneity might attribute to other factors, the insufficient data limited to identify their sources only using meta-regression.No publication bias was detected by either the inverted funnel plot or Egger’s test. The shapes of the funnel plots seemed approximately symmetrical and P values of the Egger’ tests were not statistical significant . Figure The individual susceptibility plays an important role in the development of most diseases. Therefore, the studies of genetic susceptibility are helpful for disease prevention, diagnosis, and treatment.TGFA is a well characterized mammalian growth factor. Previous studies found that although TGFA is expressed in mice during palatogenesis and mice with a null mutation of the TGFA gene have abnormal skin, hair, and eyes, but they do not have CL/P ,39. TGFAThrough this meta-analysis we found that TGFA Taq I polymorphism increased the risk of CL/P. Mitchell adopted a reappraisal and showed that there was a statistically significant association between TGFA and CL/P, However, there was evidence of significant heterogeneity from different studies, regarding an association between genetic variation at the TGFA locus and CL/P remains inconclusive . A meta-In order to reduce heterogeneity, we performed the stratified analyses. It is well know that the incidence of most genetic polymorphisms could vary between different ethnic populations and the wide range in TGFA Taq I allele frequencies across different studies suggests the heterogeneity between populations may exist . In the Despite trying our best to perform a comprehensive meta-analysis, some limitations exist in our study. Although the results for publication bias in our study were not statistically significant, our analysis used published international studies, which could arose publication bias. Lack of the original data of available studies limited our further evaluation of potential interactions, such as age, gender, family history, environmental factors and lifestyle. There were not significant results in stratified analyses among African and Hispanic population, because there were too few studies after stratifying. Therefore, more studies are needed to provide more evidence on the association between TGFA polymorphism and CL/P in different ethnic populations.Although CL/P is a complex disease, our study provided evidence showing the important role of TGFA genes polymorphisms in the development of CL/P, genetic factors determining disease susceptibility and severity may facilitate personalized medicine. Further understanding of the interactions between genetic regulatory mechanisms is critical for discovering new therapies for managing human CL/P. Specifically, targeted therapy about the polymorphisms of related genes might be a promising avenue for future CL/P diagnosis and treatment.In summary, our meta-analysis supports that the TGFA Taq I polymorphism is more likely to contribute to the susceptibility of CL/P, especially in the subgroup of Caucasian population.TGFA Taq I polymorphism may be associated with the risk of CL/P.The authors declare that they have no competing interests.CF participated in selecting the study, extracting the data, performing the statistical analysis and drafting the manuscript. EZ and WD participated in study selection, data extraction and manuscript drafting. ZX collected and extracted the data. YZ and LL participated in study selection and the statistical analysis. All authors read and approved the final manuscript.The pre-publication history for this paper can be accessed here:http://www.biomedcentral.com/1472-6831/14/88/prepub


MTHFR) gene compromises activity of the MTHFR enzyme by about 50%. The significance of correlation between 677C>T substitution and PCOS remains dubious due to the low power of published studies.Optimum efficiency of the folate pathway is considered essential for adequate ovarian function. 677 C>T substitution in the 5, 10-methylene tertrahydrofolatereductase  from India. The data analysis revealed a lack of association between this polymorphism and PCOS . Group-wise analysis on the basis of ethnicity also revealed no association in any of the ethnic groups . Homocysteine levels did not differ significantly between cases  and controls . We also undertook a meta-analysis on 960 cases and 1028 controls, which suggested a significant association of the substitution with PCOS in the dominant model of analysis (OR = 1.47 (95%CI = 1.04–2.09), P = 0.032]. Trial sequential analysis corroborated findings of the traditional meta-analysis. However, we found that the conclusions of meta-analysis were strongly influenced by studies that deviated from the Hardy Weinberg equilibrium. A careful investigation of each study and a trial sequential analysis suggested that 677 C>T substitution holds no clinical significance in PCOS in most of the populations.We analyzed MTHFR 677 C>T polymorphism does not affect PCOS risk in India. The association seen in the meta-analysis is due to an outlier study and studies showing deviation from the Hardy Weinberg equilibrium.In conclusion,  Polycystic ovarian syndrome (PCOS) is the most widespread form of female infertility. Quoted prevalence of PCOS ranges from 5 to 25% –2. The hMTHFR) an important gene for investigation in PCOS as decreased efficiency of folate/homocysteine pathway could increase the risk.Etiology of PCOS remains largely unknown; however, the syndrome is thought to be multi-factorial with hyperinsulinemia and lifestyle factors being the prominent contributing factors. Questions about the possible effect of folic acid on ovarian function were raised in the late 1960s, with the finding that deficiency/excess of folates partially inhibited ovulation in immature super-ovulated rats . Later, MTHFR gene compromises activity of this gene and the folate pathway by about 50% [It has been experimentally shown that a functional substitution, 677 C>T , in the bout 50% –16. Therbout 50% –18. On tbout 50% –20. Mostbout 50% –25. Aparbout 50% –28. We uThis study was approved by the Institutional Human Ethics Committee of the Maaruthi Medical Centre and Hospitals, Erode, Tamil Nadu and that of the CSIR-Central Drug Research Institute, Lucknow, Uttar Pardesh. We calculated the sample size using G power calculator, assuming a 10% frequency of the polymorphism , p = 0.05, study power of 90% and effect size of 0.3. This suggested a sample size of 234 (117 cases and 117 controls). We recruited 517 subjects with or without PCOS from the Maaruthi Medical Center and Hospitals, Erode, Tamil Nadu and the Institute of Medical Sciences, Banaras Hindu University, Varanasi. Relevant details about the subjects such as age, height, weight and age at marriage were noted. The inclusion criteria of patients were based on the Rotterdam Revised 2003 (2 out of 3) diagnosis guidelines. This included the presence of two out of the following three features (after the exclusion of related disorders): (1) Oligo- or an-ovulation, (2) Clinical and/or biochemical signs of hyperandrogenism, and (3) Polycystic ovaries. Hyper-androgenism was diagnosed by measuring serum testosterone, androstenedione, and DHEAS (Dehyroepiandrosteronesulphate) levels in the early follicular phase and by looking at the pattern and extent of terminal hair growth (hirsutism). Menstrual history of the patients was recorded to diagnose oligo-/ameno-rrhea. Ovarian morphology was studied by transvaginal ultrasonography to find if enlarged ovaries with at least 12 peripherally arranged immature follicles, a characteristic feature of PCOS, were seen.We recruited 261 patients (168 from the Dravidian population and 93 from the Indo-European population) following the above inclusion criteria. The exclusion criteria consisted of the presence of congenital adrenal hyperplasia, Cushing’s syndrome, thyroid dysfunction or hyperprolactinemia. The patients were categorized in obese and non-obese groups according to the World Health Organization (WHO) criteria. Women having BMI in the range of 27.3–32.3 (overweight) and >32.3 (obese) were included in the obese group while women having BMI in the range of 19.1–25.8  and 25.8–27.3  were included in the non-obese group . The BMITwo hundred fifty six ethnically matched control samples (comprising of 156 subjects from the Dravidian population and 100 from the Indo-European population) were collected from proven fertile women who volunteered to participate in the study. The controls were recruited according to the defined inclusion criteria of proven fertility with normal menstrual cycle and ovarian morphology, but no history of sub-fertility treatment. Peripheral blood samples (3–5 ml) of the patients and controls were collected for genetic analysis. Informed written consents were obtained from all participants. Homocysteine level in serum samples was measured using an enzymatic test based kit manufactured by Diazyme Laboratories (USA).Genomic DNA was isolated from peripheral blood samples according to the protocol described in our previous study . DNA conMTHFR gene was amplified using primers and the protocol published in our earlier study [2 (25 mM), 1.0 μl of dNTPs (10 mM), 2.0 pM of each of the forward (5’ CATCCCTATTGGCAGGTTACCC 3’) and reverse (5’ GGGAAGAACTCAGCGAACTCAG 3’) primers, 1.0 unit of Taq DNA polymerase enzyme (Applied Biosystems) and 40 ng of genomic DNA. PCR amplification was carried out using the ABI Veriti thermal cycler  with the following conditions; denaturation at 95°C for 10 minutes followed by 35 cycles of denaturation at 95°C for 30 seconds, annealing at 63°C for 30 seconds and polymerization at 72°C for 40 sec, and a final stage of polymerization at 72°C for 7 minutes. The amplicons were directly sequenced using BigDyeTM chain termination chemistry on the ABI 3730 DNA analyzer [677C>T region of the er study . Brieflyems, USA). Multiplwww.spss.com). The frequency of the major and minor alleles was compared between cases and controls using the chi-square (χ2) test. Pooled data for all cases (Indo-European and Dravidian) were compared with controls, followed by the comparison of cases and controls within each group. The two-sided P value of less than 0.05 (95% level of confidence) was considered significant for statistical inference.Genotype data for the control group were analyzed for fitness in the Hardy Weinberg (HW) equilibrium. All statistical analyses were performed using the SPSS software (www.pubmed.com), Google Scholar (scholar.google.co.in) and ScienceDirect (www.sciencedirect.com) databases was conducted using the keywords such as ‘PCOS’, ‘MTHFR’, ‘677C>T’ polymorphism’ and ‘polycystic ovaries’ in various combinations. The search terms were kept broad to identify all relevant articles and the last search was performed on the 28th of February 2015. The initial screening was done to look for MTHFR 677 C>T polymorphism data and the full text articles of relevant studies were obtained. The corresponding author was contacted in case a full text article was not accessible. Only fifteen studies containing data on 677 C>T polymorphism in PCOS could be identified  each trial was an independent case-control study, (ii) all the studies had a similar purpose of investigating the correlation of 677 C>T polymorphism with PCOS, (iii) enough information had been provided for the calculation of odds ratio (OR), (iv) SNP genotyping was done at the high resolution level, (v) selection of patients was done according to the standard and reliable diagnostic parameters. The studies not providing enough information (raw data) and those providing poor description of subject recruitment and analysis, were excluded.Genotype data for the target polymorphism were extracted using the standard extraction form, and wherever necessary, the genotype frequencies were re-calculated to ensure homogeneous data presentation. The data were tabulated with specific details regarding the first author, year of publication, sample size, ethnicity of study population, and allele and genotype frequencies. Details required for standard reporting in systematic reviews and meta-analysis have been presented in PRISMA checklist  and meta2) given by Higgins [Meta-analysis was conducted using the Comprehensive Meta Analysis software (version 2). We chose OR as the effect size to find PCOS risk associated with the rare genotypes (CT and TT). Raw data for each study were fed into software for calculation of the respective OR and p value. Heterogeneity test was performed using the chi-square based ‘Q’ test suggested by Cochran  and the  Higgins . In the The data from each study were tested for fitness in the Hardy-Weinberg equilibrium using the chi-square method. The impact of assumptions on the robustness of inference was judged by sensitivity analysis. Publication bias was assessed using the funnel plot of precision by log odds ratio and calculated quantitatively using various statistical tools, such as Egger’s regression intercept, Duvall and Tweedie trim and fill procedure, Classic fail-safe ‘N’ and Orwin fail-safe ‘N’. Funnel plot was generated for the picturesque analysis of publication bias.A meta-analysis is prone to systematic (bias) or random errors (play of chance) due to dispersed data and repeated significance testing. Bias from trials with low methodological quality, publication bias and small trial bias may result in false P-value. Therefore, we used trial sequential analysis (TSA) for the calculation of required information size (number of samples) for reliability of meta-analysis . Some prA comparative account of clinical data of patients falling in the hyper- and normo-androgenic groups is presented in The genotype data for the control group fitted well in the Hardy Weinberg equilibrium . No significant association between 677C>T substitution and PCOS was observed in the pooled data of all cases versus controls . Similarly, in discrete analysis for each group (Indo-European and Dravidian), the p values remained statistically non-significant for the differences between cases and controls in each group . As mentioned above, we sub-grouped Dravidian patients on the basis of BMI; however, no difference was seen in genotype comparison of obese patients versus controls (P = 0.54), non-obese patients versus controls (P = 1.0), and obese PCOS versus non-obese PCOS patients (P = 0.68).Following the above-mentioned search criteria, we could retrieve fifteen studies . Out of 2’ value of more than 50% for between studies comparison showed the presence of true heterogeneity . This further supported the choice of enrolling random effects model for the overall inference. However, the results of both the models (fixed and random) have been presented for reference.The effect sizes were spread and appeared to be the part of a normal distribution including an infinite number of studies. This favored the selection of random effects analysis model. Further, the ‘IAnalyses using both fixed effect and random effects models were undertaken with a priori preference for random effects model due to the expected heterogeneity. Analysis using the dominant model of analysis suggested a significant association of the substitution (CT + TT) with PCOS  . Outcome2 = 61.67, Tau square = 0.24, SE = 0.22, var = 0.048) and the p value turned non-significant (random effects P = 0.089) , Choi et al. (2003) and Idali et al. (2012), fitted well in the Hardy Weinberg equilibrium. Therefore, a sensitivity analysis was conducted after excluding the above three studies. The data were still significantly heterogeneous  . Therefo0 = 0.75, SE = 1.21, 95% CI = -1.98–3.48, two tailed p value = 0.55). Classic fail-safe ‘N’ value of 22 and Orwin’s fail-safe ‘N’ value of 30, suggested 2.2 and 3 studies, respectively, missing from the analysis for every study included. Since the possibility of missing so many studies is negligible, the existence of no bias seems more plausible. Picturesque view of the funnel plot also showed almost symmetrical distribution of the studies, further ruling out publication bias  [MTHFR genotypes and PCOS [Glueck et al. 1999) analyzed (P>0.09) . Later, 99 analyz(P>0.09) . Similarand PCOS . Due to and PCOS ,40.MTHFR genotype and PCOS risk in Indian populations. The two ethnically different groups provided us the opportunity to also test the effect of ethnicity on PCOS risk. The north Indian populations have an Indo-European ethnic affinity and are strikingly different from the Dravidian populations [We observed no significant correlation between the ulations . There wulations .MTHFR 677C>T polymorphism was a risk factor for PCOS in the dominant model and in CT vs TT comparison. In another meta-analysis on 607 PCOS and 663 controls from eight studies, Lee et al. (2014) reported that 677 C>T polymorphism was not a risk factor for PCOS [At least three previous meta-analyses have analyzed 677 C>T polymorphism, two of which stated no correlation , 28 whilfor PCOS . HoweverAmong the studies pooled in the analysis, only one study on a Turkish population reported a significant association of 677C>T substitution with PCOS. In order to analyze for its influence on meta-analysis, we undertook a pooled analysis after the exclusion of Karadeniz et al. 2010. The exclusion suggested no association of this polymorphism with PCOS. Therefore, 677 C>T polymorphism seems to be unrelated to PCOS in most of the populations studied so far, except in Turkish populations. This suggests the existence of sensitive studies that influenced the association of 677 C>T polymorphism with PCOS. Trial sequential analysis showed that the meta-analysis had enough sample size; however, the overall conclusion was strongly influenced by the sensitive studies. Therefore, further studies need to evaluate the association of this polymorphism with PCOS in Turkish populations. While genetic risk factors certainly affect PCOS risk, the end result may a complex outcome of an interplay between genetic and lifestyle factors such as insulin resistance, obesity and physical inactivity, which can further promote its occurrence . EventuaIn conclusion, MTHFR 677 C>T genotype does not influence the risk of PCOS in the study populations and most other populations. Further, the polymorphism is not differently distributed between obese and non-obese cases, ruling out its correlation with obesity accompanied PCOS. A direct correlation of 677 C>T polymorphism with PCOS risk was ruled out in two ethnically different Indian populations. Though meta-analysis suggested 677 C>T substitution to be a PCOS risk factor, critical analysis of the pooled data suggested that the conclusion was significantly biased. TSA validated this observation for sufficient sample size and power. Therefore, we have now reached an adequate sample size for taking a decision regarding the clinical importance of 677C>T substitution in PCOS cases. Exclusion analysis suggested that the association in the meta-analysis was largely due to the study by Karadeniz et al. (2010) on a Turkish population. The association was skewed due to the studies that deviated from the Hardy Weinberg equilibrium. Therefore, the exclusion of sensitive studies mentioned above revealed that 677 substitution has no clinical importance in PCOS. If it affects PCOS risk, the association is likely to be ethnic-specific and further studies on various populations would pave the way for identification of ethnic-specific effects. Since the association in Turkish populations is reported by one study only, further studies on Turkish or ethnically similar populations are encouraged. Studies on ethnically divergent populations such as East Asians; particularly Chinese and Japanese, may provide new data; nevertheless, there is now sufficient evidence to conclude that MTHFR 677 C>T is not clinically important in PCOS in most of the populations.S1 Table(DOC)Click here for additional data file.S2 Table(DOCX)Click here for additional data file.


TNF-β have been considered as important components in the aetiology of migraine. Many studies have investigated the association between TNF-β 252A>G polymorphism and migraine risk, with debatable results. This study was designed to examine whether the TNF-β 252A>G polymorphism confers genetic susceptibility to migraine in diverse populations.Genetic factors including TNF”, “252A>G”, “rs909253”, “polymorphism”, “polymorphisms”, “variant”, “SNP”, combined with “migraine” or “migraine with aura (MA)” or “migraine without aura (MO)”. Pooled ORs and 95% CI were appropriately calculated using the fixed-effect model.Studies eligible for this meta-analysis were searched in the PubMed, Embase, and CNKI by using the keywords “tumor necrosis factor”, “TNF-β 252A>G polymorphism and overall migraine risk. Stratified analyses by type of migraine and gender revealed similar results. Interestingly, an OR with 95% CI representing an increased migraine risk was indicated in Asians under the recessive model .We finally included a total of seven studies, providing 5 557 migraineurs and 20 543 unrelated healthy controls. Meta-analysis results showed no statistical evidence of a significant association between TNF region may modulate risk of migraine in the population of Asian ancestry.Our findings appear to support the hypothesis that genetic variability of 252A>G polymorphism in  TNF), interleukin-1β (IL-1β) and IL-10, are known genetic candidates contributing to the predisposition towards migraine development Migraine is a chronic neurological disorder accompanied with digestive system and autonomic nervous system symptoms composed of nausea, vomiting and extreme photophobia, and phonophobia TNF (TNF-α andlymphotoxin-α or LT-α or TNF-β) are cytokines implicated to influence the intensity and duration of local inflammation. TNF could act as an inflammatory mediator in the activation and sensitization of meningeal nociceptors, threshold brain excitability, and propagation of neuronal hyperexcitability, consequently leading to persistent pulsating headache characterized in migraine TNF-β mapped on chromosome 6 is a member of the TNF cytokine superfamily, playing an important role in immune and inflammatory responses TNF-βTNF-β region is linked with the risk of migraine. We hypothesized that the 252A>G polymorphism may be associated with migraine risk. The hypothesis was examined though a meta-analysis of 5 557 migraineurs and 20 543 unrelated healthy controls from seven publications.The aim of this study was to determine whether 252A>G polymorphism in the TNF-β 252A>G polymorphism and migraine were systematically searched using the keywords “tumor necrosis factor”, “TNF”, “252A>G”, “rs909253”, “polymorphism”, “polymorphisms”, “variant”, “SNP”, combined with “migraine”, or “migraine with aura (MA)” or “migraine without aura (MO)” in the PubMed, Embase, and China national knowledge infrastructure (CNKI) by two independent investigators. A manual search for potentially relevant studies was also conducted by reviewing the literature identified in the electronic databases. The last search was updated in June, 2013.Studies examining the association between TNF-β 252A>G polymorphism and migraine risk between patients and matched controls; ii) that contained genotype frequency in full detail allowing for calculation of crude risks for migraine. We excluded the studies reporting the association between TNF-β 252A>G polymorphism and survival from migraine and/or response to any therapy. If more than one article was published by the same author sharing the same cases series, we selected the study with the most subjects.The meta-analysis only included the articles i) that evaluated the association of Usable data were independently extracted from each paper by two investigators using a structured sheet and then pooled into a database. The items recorded were first author, publication year, location of the study, racial descent of the studied subjects, type of controls, mean age of the cases, total number of cases and controls for migraine, MA and MO, and the corresponding genotype distributions.2 test for each control group of the included studies. An odd ratio (OR) with 95% confidence interval (CI) was measured to estimate the association between 252A>G polymorphism and migraine risk. The significance for pooled ORs was evaluated by Z test. The ORs of homozygous model (GG vs. AA), dominant model (GG+AG vs. AA), recessive model (GG vs. AG+AA), allele model (G vs. A), and heterozygous model (AG vs. AA) were calculated by the fixed effects model or the random effects models according to the outcome of between-study heterogeneity detected by the Chi square based Q test Statistic analysis was performed by using STATA version 12.0 . The significance threshold was defined as P<0.05 unless specially stated. Deviation from Hardy-Weinberg equilibrium (HWE) was tested using the goodness-of-fit χStratified analyses were performed by racial descent, type of migraine, and gender. The leave-one-out sensitivity analysis was carried out to detect the extent of influence from the single studies on the combined results. The funnel plot and the Egger's test were applied to evaluate publication bias across studies The systematically electronic along with manual searches supplied 137 papers that may have usable data for this meta-analysis. However, 126 of them were removed after preliminary review of titles and abstracts. Among the remaining 11 papers, we screened the full texts to examine their eligibility based on the inclusion criteria. As a result, 4 studies were excluded  When stratifying the data by type of migraine, we did not observe either an increased risk or decreased risk of MA or MO. Subgroup analysis based on gender provided similar results that as compared with men, women had no higher risk of developing migraine or MO .To explore the potential sources of heterogeneity on the overall results, we conducted sensitivity analysis and the initial ORs were not significantly influenced by sequential removal of each study from the total analysis. Egger's test and the Begg's funnel plots performed in all contrast models provided no statistical evidence of publication bias  .TNF is a cytokine involved in the mediation of inflammatory reactions and endothelial function, and it has been implicated in inflammation-related hyperalgesia TNF-α, a potential pain mediator in neurovascular inflammation, has been reported to increase migraine risk in the population of Italian ancestry TNF-β is another subunit of TNF gene and has been frequently investigated in the studies concerning migraine risk.Migraine is a primary headache disorder with its aetiology being incompletely understood. In recent years, a great deal of attention has been directed to the field of migraine genetics TNF gene polymorphisms and migraine presented the discoveries indicating that there was no overall association between the 252A>G polymorphism and migraine. Neither did the investigators find effect modification of MA or MO in women or men in the meta-analysis, where ethnicity-specific analysis was not carried out A previous investigation into the associations between In our study, we observed statistical evidence for a significant association of the 252A>G polymorphism with migraine in Asian subjects, but not in Caucasians. Multiple possibilities may result in the differential relation, but there are two most likely explanations. One refers to the study size. Notably, the number of Caucasian subjects is much larger relative to the Asian subjects (24 637 vs. 1 463), which assures an adequate detection power in Caucasian population. And we can not exclude the possibility that the current positive association revealed in Asians may be derived by chance. The other explanation may be the different frequency of the G allele between the two populations: for Asians, the frequency is 37.4% and 22.6% for Caucasians. This implies that the genetic background is a potential factor influencing the genetic predisposition to migraine.Heterogeneity and publication bias are two inevitable confounding factors which could distort the true associations between SNPs and diseases under research when performing meta-analysis. Considerable attention should be paid in interpreting the findings if the two factors are obviously indicated in related tests. Fortunately, as no significant heterogeneity and publication bias were revealed in our meta-analysis, the results for the association assessed in the present study are relatively convincing by using a rigorous analytic approach. However, the possibility that the effect size of risk estimate is derived by chance can not be excluded, because the total number of cases in general analysis, not to mention the subgroup analyses, is not large enough to obtain a precise assessment of the association between 252A>G polymorphism and migraine. Further, influence from environmental factors is of substantial importance in risk estimate. Such exploration was precluded owing to the inadequate data.To sum up, these data provided further evidence that 252A>G polymorphism in the TNF region may represent a risk factor for the development of migraine in the population of Asian ancestry. No statistical evidence was indicated in subgroup analyses by type of migraine and gender. Further investigations using a much larger sample with interplay between environment and gene taken into account are clearly needed to validate the findings in this meta-analysis.Checklist S1PRISMA Checklist.(DOC)Click here for additional data file.


SHMT1 C1420T may lead to the abnormal biosynthesis involved in DNA synthesis and methylation, and it may eventually increase cancer susceptibility. Many epidemiologic studies have explored the association between C1420T polymorphism and the risk of non-Hodgkin lymphoma (NHL), but the results have been contradictory. Therefore, we performed this meta-analysis to evaluate the relationship.Serine hydroxymethyltransferase 1 (SHMT1) is a key enzyme in the folate metabolic pathway that plays an important role in biosynthesis by providing one carbon unit. SHMT1 C1420T polymorphism on NHL risk. Odds ratios (ORs) and 95% confidence intervals (CIs) were calculated to measure the strength of the association.The meta-analyses were conducted to evaluate the effect of SHMT1 C1420T polymorphism and NHL risk under the allelic comparison ; a borderline association was found between SHMT1 C1420T polymorphism and NHL risk under the homozygote model  and the dominant model .Eight studies encompassing 3232 cases and 4077 controls were included. A statistically significant association was found between SHMT1 C1420T polymorphism may be associated with NHL risk, which needs to be validated in large, prospective studies. In the past 30 years, the incidence of non-Hodgkin lymphoma (NHL), a common hematologic malignancy, has increased markedly , 2. GeneSHMT1, localized on chromosome 17p11.2 [SHMT1 can affect cell progression and ultimately cause cancer. One SNP has been found at nucleotide 1420  [SHMT1 by a leucine-to-phenylalanine amino acid substitution at codon 474 (Leu474Phe) of the protein [Serine hydroxymethyltransferase (SHMT), a key enzyme involved in the folate metabolism, can reversibly catalyze serine and tetrahydrofolate to glycine and 5,10-methylenetetrahydrofolate [ 17p11.2 . Abnorma 17p11.2 , 14, whi 17p11.2 . Consequ1979277) , and it SHMT1 C1420T polymorphism and NHL risk, but the conclusions are mixed rather than conclusive, partially because of possible weak effects of the polymorphisms on NHL risk, the relatively small sample size in each previous investigation, or the patients’ diverse racial backgrounds. Therefore, we developed a comprehensive meta-analysis of all eligible case–control studies to derive a more precise risk estimate for the association.To date, many studies have investigated the association between SHMT”, “SHMT1”, or “cytosolic serine hydroxymethyltransferase”, “polymorphism” or “variant”, “non-Hodgkin lymphoma” or “NHL”, and “cancer”, “neoplasia”, or “malignancy” . To expand the scope of our search, we also searched the Chinese National Knowledge Infrastructure database (CNKI) with the terms “SHMT”, “SHMT1”, and ‘‘NHL’’ in Chinese. Furthermore, we manually searched reference lists on this topic to identify additional relevant studies and attempted to contact the authors for more information if the information available was incomplete.We searched two electronic databases (PubMed and Embase) to identify all published studies with the following terms: “SHMT1 C1420T polymorphism and NHL risk; and (4) provide sufficient data for the calculation of odds ratios (ORs) and 95% confidence intervals (CIs). Abstracts and unpublished reports were excluded. Moreover, if studies had the same subjects or overlapping data, we selected the one with the largest sample size.The studies selected in this meta-analysis had to meet the following criteria: (1) be written in English or Chinese; (2) have a case–control design; (3) evaluate the association between Two reviewers independently abstracted the following information from each study according to standardized criteria: first author, year of publication, country of population studied, ethnicity of population studied, NHL subtype (DLBCL or FL), source of controls , total number of genotyped cases and controls, and numbers of genotypes  for the C1420T polymorphism in cases and controls. If any different views existing, we discussed it until consensus was reached.SHMT1 C1420T polymorphism and NHL risk under different genetic models as follows: allelic comparison (T vs. C), homozygote model (TT vs. CC), heterozygote model (CT vs. CC), recessive model (TT vs. CC+CT), and dominant model (CT+TT vs. CC). Additionally, we performed stratification analyses by tumor subtype (DLBCL and FL) and by patient ethnicity . Goodness-of-fit Chi-square test was used to evaluate deviation from the Hardy–Weinberg equilibrium (HWE) for the genotypes of controls. The Chi-square-based Q test was performed to calculate inter-study heterogeneity. If P < 0.05, we used the random-effects model to assess the pooled ORs because this model tends to provide wider 95% CIs [P < 0.05 as determined by the Egger test was considered statistically significant [P < 0.05 was considered statistically significant.We used the crude ORs and 95% CIs to determine the association between  95% CIs ; otherwi 95% CIs . FurtherSHMT1 polymorphism, and 1 was in a language other than English or Chinese), and 9 articles were left for full review. Of these, 1 study was excluded for not providing genotype frequencies of NHL [SHMT1 C1420T polymorphism and the subtype of NHL risk [After initial screening, we found 47 relevant publications. We excluded 38 of these studies after reviewing the titles and abstracts ; we found a borderline association between SHMT1 C1420T polymorphism and NHL risk under the homozygote model  and under the dominant model . In the subgroup analysis by ethnicity, we found no association for Caucasians but a borderline direct association for mixed ethnic subgroups under the allelic comparison  and the dominant model .The primary results of this meta-analysis are presented in Table After excluding the study by Li et al.  because P value. We further used the false-positive report probability (FPRP) with prior probabilities of 0.0001, 0.001, 0.01, 0.1, and 0.25 to account for chance associations from multiple comparisons , except for the recessive model (P = 0.04). In the sensitivity analyses, the results indicated that a single study might change the pooled ORs (data not shown).No significant between-study heterogeneities were observed among the overall studies for the association of P = 0.83; homozygote model: P = 0.39; dominant model: P = 0.29; recessive model: P = 0.23), but the heterozygote model showed significant publication bias (P = 0.02).Begg’s funnel plot and the Egger test were used to evaluate the publication bias of all included studies. The shapes of the funnel plots appeared to be symmetrical, and the Egger test further suggested that there was no significant evidence of publication bias under some genetic models  could not be obtained for stratification analysis to further evaluate the relationship between SHMT1 C1420T polymorphism may be a risk factor for NHL. Additional well-designed studies with larger sample sizes and more information about confounding factors are needed to validate our findings.In summary, we found in this updated meta-analysis that


Accumulating evidence supports a role of DNA methylation in the pathogenesis of leukemia. The aim of our study was to evaluate the potential genes with aberrant DNA methylation in the prediction of leukemia risk by a comprehensive meta-analysis of the published data.A series of meta-analyses were done among the eligible studies that were harvested after a careful filtration of the searching results from PubMed literature database. Mantel-Haenszel odds ratios and 95% confidence intervals were computed for each methylation event assuming the appropriate model.CDKN2A, CDKN2B, ID4 genes were significantly hypermethylated in acute myeloid leukemia.A total of 535 publications were initially retrieved from PubMed literature database. After a three-step filtration, we harvested 41 case-control articles that studied the role of gene methylation in the prediction of leukemia risk. Among the involving 30 genes, 20 genes were shown to be aberrantly methylated in the leukemia patients. A further subgroup meta-analysis by subtype of leukemia showed that Our meta-analyses identified strong associations between a number of genes with aberrant DNA methylation and leukemia. Further studies should be required to confirm the results in the future. Recent advances attest to the great promise of DNA methylation markers as powerful tools in the clinic A systematic literature search was performed in PubMed by using “leukemia” and “DNA methylation” as the search terms for articles updated until December 25, 2013. The search was limited to articles published in English and Chinese. Articles in the search output were surveyed according to their titles and abstracts. Studies were selected if they met the following criteria: 1) they were case-control associations of gene methylation with the risk of leukemia in humans; 2) they had sufficient methylation information to calculate the odd ratios (ORs) and 95% confidential intervals (CIs) for the meta-analysis. The selection procedures of studies were illustrated in the flow chart of For the eligible articles, we extracted the following information: first author’s name, published year, PubMed ID, disease category , the numbers of cases and controls. All the data were extracted by four authors . A consensus was reached through a rigorous discussion when there existed conflicting evaluations.2 metric 2>50%), otherwise a fixed-effect model was applied for the meta-analysis Review manager 5 was used for meta-analysis. Mantel-Haenszel ORs and their 95% CIs were computed for each gene to evaluate the contribution of gene methylation to the risk of leukemia. Heterogeneity of the studies in the meta-analysis was evaluated by IA total of 535 studies were retrieved from the PubMed literature database after searching the keywords of “leukemia” and “DNA methylation”. After a series of selection procedure shown in CDKN2A methylation was involved with 12 case-control studies among 576 controls and 167 cases. Our results showed a significant heterogeneity among the 12 studies (I2 = 65%). In addition, the meta-analysis indicated that hypermethylation of CDKN2A gene was associated with the increased risk of leukemia . Meta-analysis between GLIPR1 methylation and leukemia was involved with 6 case-control studies among 384 controls and 309 cases (2 = 83%) among these studies, and revealed that GLIPR1 hypermethylation was associated with the increased risk of leukemia . Meta-analysis of CDKN2B methylation with leukemia was carried out in 5 studies among 86 controls and 108 cases (2 = 56%), and revealed that hypermethylation of CDKN2B gene was associated with the increased risk of leukemia .As shown in 09 cases . The met08 cases . Our resSHP1 methylation was involved with 5 case-control studies among 46 controls and 104 cases. Our results indicated that hypermethylation of SHP1 gene was associated with the increased risk of leukemia . Meta-analysis between ID4 methylation and leukemia was involved with 5 case-control studies among 78 controls and 165 cases. The results revealed that ID4 hypermethylation was associated with the increased risk of leukemia . Meta-analysis of DAPK1 methylation with leukemia was carried out in 3 studies among 29 controls and 169 cases. Our results showed that hypermethylation of DAPK1 gene was associated with the increased risk of leukemia . In addition, our results were unable to observe significant association of the methylation of CDKN1C and RARA genes with leukemia.Meta-analysis of CDKN2A methylation in AML was involved with 5 case-control studies among 80 cases and 224 controls. A significant heterogeneity was observed among the 5 studies (I2 = 72%). Our results showed that CDKN2A hypermethylation was associated with the increased risk of AML . The meta-analysis between CDKN2B methylation and AML was involved with 3 studies among 70 cases and 50 controls (CDKN2B methylation (I2 = 76%). Our results showed CDKN2B hypermethylation was associated with the increased risk of AML . As shown in ID4 methylation was involved with 3 studies among 103 cases and 48 controls. The results showed that ID4 hypermethylation was associated with the increased risk of AML .We further performed a breakdown meta-analysis by leukemia subtypes in the above-mentioned studies . As showcontrols . There eNumerous studies have found that DNA methylation of different genes was associated with the risk of leukemia CDKN2A and CDKN2BCDKN2B and CDKN2A CpG islands is frequent in malignant transformation CDKN2A gene and CDKN2B gene were risk factors for leukemia, especially for AML. In the subgroup analysis, we found that DNA methylation of CDKN2A gene was significantly associated with AML, but not with CML or CLL. A microarray analysis in 2011 identified that glioma pathogenesis-related protein 1 (GLIPR1) was a methylation-silenced gene in the AML patients, and might serve as a marker to monitor the therapeutic effect of AML GLIPR1 gene was a risk factor for leukemia. GLIPR1 is a pleiotropic protein involved in cell proliferation, tumor growth and apoptosis, and it may affect G protein signaling and cell cycle regulation SOCS1 is an important protein in the JAK/STAT pathway, and plays a key role in the downstream regulation of BCR-ABL protein kinase SOCS1 gene was a protective factor for leukemia. SHP1 is tumor suppressor gene involved in the regulation of cell cycle control and apoptosis SHP1 is negative regulator of the Jak/STAT signaling pathway that is implicated in leukemogenesis. Promoter methylation of SHP1 gene is able to silence its gene expression, and was frequently detected in various kinds of leukemias and lymphoma SHP1 gene was a risk factor for leukemia. Inhibitor of DNA binding protein 4 (ID4) is a member of the dominant-negative basic helix-loop-helix transcription factor family that lacks DNA binding activity and has tumor suppressor function. Promoter of ID4 is consistently methylated to various degrees in CLL cells, and increased promoter methylation in a univariable analysis was shown to be correlated with shortened patient survival ID4 gene was a risk factor for leukemia, especially for AML. Death-associated protein kinase 1 (DAPK1), a tumor suppressor, is a rate-limiting effecter in an endoplasmic reticulum stress-dependent apoptotic pathway DAPK1 have been demonstrated to be key pathogenic events in CLL DAPK1 gene was a risk factor for leukemia.De novo methylation of the 5′CpG island has been reported as an alternative mechanism of inactivation for tumor suppressor genes The current meta-analysis has some limitations. Firstly, selection bias is inevitable due to the search strategy restricted to articles published in English or Chinese. Secondly, some large between-study heterogeneity existed in our meta-analyses. This phenomenon may be caused by the facts that different subtypes of leukemia were not separated in the first place due to the limited studies, and that different regions of the same gene were tested for methylation in the involved studies. Thirdly, this analysis was performed at the study level, which limited ability to explore the potential for confounding by various demographic and clinical factors . Fourthly, most of the studies we selected were performed with Methylation-Specific PCR (MSP) and the status of DNA methylation was qualitative (M+ or M−), and it also limited the scope of our analysis. Finally, the efficiency and accuracy of statistical analysis may be influenced by the moderate amount of subjects. We expect more samples being tested in the future to draw a more reliable conclusion.In conclusion, the results of this study indicated that certain genes DNA methylation was independently associated with the risk of leukemia, especially some kinds of leukemia. Also more studies should be required to confirm the results in the future. DNA methylation has a very strong potential to be a useful biomarker for predicting, prognostication and prediction of response to chemotherapy of leukemia.Checklist S1PRISMA Checklist.(DOC)Click here for additional data file.


A quantity of case-control studies have been performed to address the association between three cyclooxygenase-2(COX-2) polymorphisms  and the risk of hepatocellular carcinoma (HCC). However, previous research results are inconsistent. We conducted this meta-analysis to clarify the correlation between these COX-2 polymorphisms and HCC risk.The authors searched in PubMed, EMBASE, Google Scholar, CNKI and WanFang database for relevant articles up to April 28, 2014. The data were extracted by two independent reviewers. Odds ratios (ORs) and 95% confidence intervals were calculated.A total of 8 studies consisting of 2182 cases and 3324 controls were included in this meta-analysis. For COX-2 polymorphism -1195G/A, an association with increased risk was observed under the heterogeneous, homozygous, dominant model. However, COX-2 polymorphisms (-765G/C and +8473T/C) were not related to HCC risk in this study. We also found a similar result in the subgroup analysis of Chinese population that -1195G/A polymorphism, instead of -765G/C or +8473T/C polymorphism, was correlated with the risk of HCC.Polymorphism -1195G/A of COX-2 might be associated with susceptibility to HCC, but no similar correlations were observed between polymorphisms (-765G/C and +8473T/C) and HCC risk. Further large and well-designed studies are required to validate this association. Hepatocellular carcinoma (HCC) is the sixth most prevalent cancer and the third cause of cancer-related death in the world, which continues to be a significant public health problem . The devin vitro and in vivo [The cyclooxygenase-2 (COX-2) is one of the key enzymes of prostaglandin pathway, since it can convert the arachidonic acid to prostaglandins. Increasing evidence points to COX-2, which contributes to immune evasion, angiogenesis regulation and apoptosis inhibition, as a risk factor in carcinogenesis of HCC , 5. Some in vivo –8. Mecha in vivo . Moreove in vivo –18, but  in vivo , 15, 18, in vivo , 17. As  in vivo . Hence, We performed a systemic search in PubMed, EMBASE and Google Scholar as well as Chinese databases including China National Knowledge Infrastructure (CNKI) and WanFang database for all the relevant studies utilizing the following search terms: “cyclooxygenase-2” or “COX-2”, “polymorphism” or “SNPs”, “hepatocellular carcinoma” or “liver cancer” . Additionally, articles in the reference list were manually searched for potentially relevant studies. Publication language was not restricted. Search and literature retrieval were completed independently by two of the authors (Zhigang Chen and Jiye Zhu). Disagreements of the search result were settled by discussion among all the authors.Studies were included in the meta-analysis if they met the following criteria: (1) the study assessed the association between hepatocellular carcinoma risk and COX-2 polymorphisms  (2) the diagnosis of HCC was validated histologically or pathologically. (3) case-control studies (4) data in the studies were adequate to calculate odds ratio (OR) and 95% confidence interval (95%CI). In the condition of multiple studies based on the same or overlapping population from the same research team, the study providing largest population was included. The main criteria for exclusion of publications: (1) family-based or sibling-based association studies (2) the study without control group. (3) duplicated publication (4) master theses and doctor dissertations (5) literature with insufficient data for evaluating OR and 95%CI.Two authors (Zhigang Chen and Jiye Zhu) independently extracted data conforming to the inclusion and exclusion criteria from the eligible articles. Any discrepancies in the process were resolved by consensus. Certain data were retrieved from the included studies: the first author’s name, publication time, ethnicity of the research population, the method of detecting genotype, the number of cases and controls, the genotyping distribution of cases and controls, polymorphism site and the Hardy-Weinberg equilibrium (HWE) results.2 statistic and the presence of heterogeneity was confirmed if the result was PQ < 0.05 or I2≥ 50%. In the condition of existence of heterogeneity, a random-effect model was utilized [We measure the strength of the association between COX-2 SNPs  and HCC risk by pooled OR with its corresponding 95% CI. The pooled ORs were estimated for allelic comparison , homozygote comparison and heterozygote comparison, dominant model  and recessive model , respectively. The P value of the pooled OR was considered significant if less than 0.05, which was examined by Z test. Heterogeneity across studies was determined by Chi-square test based Q statistic test and Iutilized , 21; othutilized . AdditioAs shown in the PA = 0.026; AA vs GG: OR = 1.466, 95%CL 1.194–1.801, PA = 0.000; GA+AA vs GG: OR = 1.356, 95%CL 1.148–1.602, PA = 0.000, Fig. 2). However, there was no association in allelic comparison and recessive model . Under the dominant model, recessive model as well as other three comparisons, no evidence supported obvious correlations between COX-2+8374T/C and -765G/C and HCC risk .Table 3) indicated that -1195G/A polymorphisms of COX-2 was remarkably associated with HCC risk in Chinese population . It seemed that COX-2-765G/C was not correlated with susceptibility to HCC in Chinese population . As only one or two studies of the Turkish or Egyptian were included, the analysis result tended to be less reliable, of which detailed data were not showed in this study.Subgroup analysis by ethnicity . Besides, it also appeared to be associated with HCC risk in the heterogeneous comparison . This result is similar to a previous meta-analysis including merely 5 articles [ours = 0.000, P et al.Bu, X.,  = 0.001; dominant model: Pours = 0.000, P et al.Bu, X.,  = 0.011)[A = 0.001, GA+AA vs GG:OR = 1.290, 95%CL 1.069–1.555, PA = 0.008).Thus, we can deduce that COX2-1195G/A tends to facilitate the development of HCC to some extent, but this result was still required further studies to confirm the correlation with HCC risk.In our meta-analysis, relationships between three COX-2 polymorphisms have been investigated. For-1195G/A, it has been fully discussed in a previous meta-analysis that aims to clarify the relationship between COX-2-1195G/A and digestive system cancers, which, however, seems not to include HCC . Therefoarticles , which r = 0.011). The low = 0.011). Aiming The COX2-765G/C has been reported to be strongly associated with skin, nasopharyngeal, gastric carcinoma –28. In aThe COX-2 +8473T/C is relevant to oral squamous cell carcinoma and breast malignancy and a study showed that the COX-2 +8473T/C polymorphism is not associated with lung cancer risk proved by meta-analysis , 32. ThrBefore we made a deduction from a meta-analysis, a number of limitations and shortcomings should be considered. Firstly, No certain studies using a prospective design confirm the correlation between three COX-2 polymorphisms  and HCC risk. None of the studies included in our meta-analysis is not retrospective hospital-based case-control study and thus a prospective study should be applied to study the causal relationship between three COX-2 polymorphisms  and HCC risk. Secondly, the quantity of studies and cases involved are limited, especially in the analysis of COX-2-765G/C and +8473T/C. To be more specific, only 5 studies about COX-2-765G/C and 3 studies about COX-2+8473T/C were included. The majority of articles in this study were about Chinese and Egyptian and no studies of other races were included in this meta-analysis. The meta-analysis suggested the possible association between COX-2-1195G/A and HCC risk, whilst the correlation in other ethnic groups remain ambiguous as before. Hence large sample size studies targeting to other racial population ought to be designed and conducted in the near future. Finally, some environmental risk factors like HBV/HCV infection, liver cirrhosis and alcohol intake were not considered in this meta-analysis as a result of insufficient data. It is likely that those risk factors may be related to the association between COX-2-1195G/A and HCC risk, but no evidence relevant to the possible interaction was mentioned in the included studies. Apparently, further well-designed studies in which other risk factors are thoroughly considered are essential for casting light on relationships among environmental risk factors, COX-2 polymorphisms and HCC.In conclusion, the COX-2-1195G/A polymorphism might have an association with HCC risk, but COX-2-765G/C and COX-2+8473T/C are not likely to exert any influence on the susceptibility to HCC. However, data of this meta-analysis were restricted to Chinese, Turkish and Egyptian population and no publications involving other ethnic groups were included. Large samples size and well-designed studies are urgently needed.S1 Checklist(ZIP)Click here for additional data file.S1 Support Information(ZIP)Click here for additional data file.S2 Support Information(DOCX)Click here for additional data file.