PIPN) cannot be predicted from clinical parameters and might have a pharmacogenomic basis. Previous studies identified single nucleotide variants (SNV) associated with PIPN. However, only a subset of findings has been confirmed to date in more than one study, suggesting a need for further re‐testing and validation in additional clinical cohorts. Candidate PIPN‐associated SNVs were identified from the literature. SNVs were retested in 119 patients selected by extreme phenotyping from 269 in NCCTG N08C1 (Alliance) as previously reported. SNV genotyping was performed by a combination of short‐read sequencing analysis and Taqman PCR. These 22 candidate PIPN SNVs were genotyped. Two of these, rs7349683 in the EPHA5 and rs3213619 in ABCB1 were found to be significantly associated with PIPN with an Odds ratios OR = 2.07 (P = 0.02) and OR = 0.12 (P = 0.03), respectively. In addition, three SNVs showed a trend toward a risk‐ or protective effect that was consistent with previous reports. The rs10509681 and rs11572080 in the gene CYP2C8*3 showed risk effect with an OR = 1.49 and rs1056836 in CYP1B1 showed a protective effect with an OR = 0.66. None of the other results supported the previously reported associations, including some SNVs displaying an opposite direction of effect from previous reports, including rs1058930 in CYP2C8, rs17222723 and rs8187710 in ABCC2, rs10771973 in FGD4, rs16916932 in CACNB2 and rs16948748 in PITPNA. Alliance N08C1 validated or supported a minority of previously reported SNV‐PIPN associations. Associations previously reported by multiple studies appeared to have a higher likelihood to be validated by Alliance N08C1.Paclitaxel‐induced peripheral neuropathy ( Chemotherapy‐induced peripheral neuropathy (CIPN) is a toxicity that adversely affects a substantial minority of patients treated with paclitaxel, while others receiving the same drug remain unaffected CYP2C8*3 was reported and validated by two independent groups in three different patient cohorts FZD3, rs7349683 in EPHA5, rs4737264 in XKR, and rs10771973 in FGD4 as candidate PIPN biomarkers EPHA5) and rs4737264 (XKR).Of the reported SNV, only a subset has been included in more than one study and of these only a few have been validated across multiple patient cohorts. An association of rs10509681 in the drug metabolism gene ABCB1) were each proposed by one report and subsequently refuted in follow‐up studies by others However, several associations of SNV with PIPN, including some of those introduced above, could not be confirmed by other studies. Rizzo et al. and Ofverholm et al. failed to confirm an association of rs10509681 (CYP2C8*3) with PIPN ABCB1) along with several other candidates LIMK2 as a candidate biomarker of PIPN, which did not reach the threshold for genome‐wide significance in their report suggesting a need for future independent validation.Recently, Abraham et al. reported novel PIPN risk SNV rs3213619  N08C1 A systematic review of the literature was performed to identify previously reported studies in PIPN pharmacogenomics. A MEDLINE/PubMed search with the key words “Chemotherapy induced peripheral neuropathy,” or “paclitaxel neuropathy,” or “taxane pharmacogenetics,” or “paclitaxel neurotoxicity,” was carried out in October 2014. The bibliographies of the identified publications were also reviewed for additional studies. Only those studies that tested direct associations between SNV and PIPN as one of their endpoints were included. Studies that investigated SNV associations with other outcomes were excluded. Only reports in the English language published in peer‐reviewed journals were considered. In the case of publications reporting data on multiple SNV, only positive results were selected for retesting based on the strength of the association reported or its location in a gene with strong evidence for an association with PIPN. All SNV fulfilling the above criteria that were also located in transcribed regions of a gene, that is, exonic or adjacent untranslated regions (UTR) were included in the present study. In addition, 9 intronic SNVs were selected  N08C1 is a previously reported prospective PIPN clinical trial that was designed to study the natural history of paclitaxel neuropathy FZD3), rs7001034 (FZD3), rs2233335 (NDRG1), rs10771973 (FGD4), rs16948748 (PITPNA), rs17781082 (CAND1), rs16916932 (CACNB2), rs4737264 (XKR4), rs1903216 (BCL6) were genotyped by TaqMan® PCR. PCR amplification of gDNA encompassing the loci of interest was performed to generate DNA amplicons. TaqMan probes directed to the allele at the test loci were designed. Analysis of the fluorescence signal from the Taqman PCR reaction and allele discrimination was performed with the SDS 2.0 software . gDNA for Taqman PCR genotyping was available for 114 (of 119) patients (71 cases and 43 controls). Experiments were performed at Mayo Clinic, Rochester, MN.Genomic DNA (gDNA) was isolated from peripheral leukocytes from each patient. Quality control of the DNA was done with Qubit®  fluorometer and NanoDrop®  spectrophotometer. Thirteen SNVs in 8 genes were genotyped by the sequencing‐based approach described previously R. Fisher exact test P‐value of <0.05 was considered statistically significant. Collection of the clinical data was conducted by the Alliance Statistics and Data Center. All analyses were based on the study database frozen on 17 December 2012.Statistical analysis was done under the three models of inheritance, additive , dominant, and recessive. Odds ratios (OR) were computed with their corresponding 95% confidence intervals (CIs) using the statistical programming package The 22 SNVs in 16 genes were genotyped. The minor allele frequencies (MAFs) in the present study were similar to the original report or, in cases of divergence, tended to be in‐between the original report and the MAF found in the reference dataset dbSNP  and the same direction of effect as the original study ABCB1 had an OR = 0.12 (P = 0.03) with the minor allele being protective, as in the original report Two SNVs were found to be significantly associated with PIPN, rs7349683 in the gene P < 0.05 in the present study). The two SNVs rs10509681 and rs11572080 in the gene CYP2C8*3, which are in perfect linkage disequilibrium (LD) (R2 = 1), showed a PIPN risk phenotype with an OR = 1.49 for the additive model and OR = 1.56 for the dominant model in the present study. This compared with previous studies having reported a range of findings, which included strong effects with an OR = 3.13 (dominant model reported only) CYP2C8*3 SNV with PIPN CYP1B1 showed a protective effect with an OR = 0.66 in the present study, which was a similar effect size and in the same direction as the original report Of the remaining SNVs, some showed a trend toward a risk or protective effect in the same direction as the original report without meeting the cutoff for statistical significance (of FGD3 had an OR = 0.9 compared with OR = 0.57 in the previous report CYP2C8 had an OR = 0.26 (P = 0.05) in contrast to the original study by Abraham et al. that had a reported risk effect with an OR = 1.48 CACNB2 had an OR = 0.29 (P = 0.08) in the present cohort, while an OR = 2.08 had been observed in the prior study ABCC2 that are in strong LD had an OR = 1.94 and 1.97, respectively, in contrast to the previously published reports that showed a decreased risk of PIPN with an OR = 0.66 and OR = 0.63 FGD4 had an OR = 0.56 and rs16948748 in PITPNA had an OR = 0.60 also trending in the opposite direction of the previous report, where the HR was 1.57 and 2.37, respectively The remaining SNV had an OR of 0.67–1.5, considered here to be a negligible effect (close to OR = 1), or had an opposite direction of effect from the original report. Eleven SNVs had an OR closer to 1 which include rs7001034 in ABCB1), rs1903216 (BCL6), rs2233335 (NDRG1), rs7001034 (FZD3), and rs4737264 (XKR4), the CI around the OR in the present study encompassed the OR in the original report, suggesting that the power of the present study was not sufficient to draw conclusions. For the rs17781082 (CAND1), and rs7833751 (FZD3) the previously reported OR was outside of the CI for the present study, suggesting that the new cohort provided contradicting evidence. (For rs10932374 (ERBB4), rs4141404 (LIMK2), and rs8110536 (MISP/C19orf21) HR instead of OR were provided in the original studies.)The 11 SNVs that were contradicting or not supportive of prior reports and considered here as negative had MAF in the present study that were similar (within 1.5‐fold) to the original report and to reference datasets, and none diverged significantly from the allele distribution expected on the basis of a Hardy–Weinberg equilibrium. For the SNV rs1045642 and rs2032582  P = 1.4 × 10−9 on meta‐analysis) ABCB1. This SNV was significantly associated with reduction of risk of PIPN in the study by Abraham et al. which reported an OR = 0.47  in a study of 1303 patients. The present study replicated the finding (P = 0.03) CYP2C8*3 was associated with risk of neuropathy without reaching statistical significance. The association of variant in CYP2C8*3 was first reported by Green et al. N08C1 is a recently reported new PIPN cohort that differs from the previously performed PIPN pharmacogenomics studies by its extensive serial PIPN phenotyping with the CIPN20 instrument. N08C1 is therefore a suitable cohort for attempting to validate previous observations from other studies. N08C1 represents a relatively small cohort of only 119 patients. The limited cohort size is in part due to the rigorous patient selection criteria and extreme phenotyping approach used; these factors at the same time mitigate the impact of limited study size by using the patients that can provide the strongest signal. Study size impacts the confidence of the results differently for each SNV depending on its MAF. We addressed this by reporting the statistical CI for each OR in Table EPHA5 belongs to the ephrin family of tyrosine kinase receptors involved in communication and signaling between different cell types in the nervous system EPHA5 signaling plays an important role in early stages of synaptogenesis ABCB1, also known as MDR1 (multidrug resistant gene), is a part of the ABC transporter superfamily. It encodes a P‐glycoprotein (Pgp), which was first described in multi‐drug resistant cells CYP2C8 and CYP3A4 are involved in the metabolism of paclitaxel. Variants in gene CYP2C8 decrease paclitaxel metabolic activity and lead to increased drug exposure The SNVs validated in the present study are found in genes involved in peripheral nervous system biology or in drug metabolism. The gene Taken together, these results indicate that SNV associations with CIPN should be supported by multiple independent studies before any consideration as a clinical biomarker.The authors declare no conflict of interest.


Epidemics of hand, foot and mouth disease (HFMD) among children have caused concern in China since 2007. We have conducted a retrospective study to investigate risk factors associated with HFMD. In this non-matching case-control study, 99 HFMD patients and 126 control from Guangdong Province were enlisted as participants. Data comprising demographic, socio-economic, clinical and behavior factors were collected from children’s parents through face-to-face interviews by trained interviewers using a standardized questionnaire. Results of the primary logistic regression analyses revealed that age, history of cold food consumption, hand-washing routines, and airing out bedding were significantly associated with HFMD cases. Results of further multivariate analysis indicated that older age  and hand-washing before meals  are protective factors, whereas airing out bedding more than thrice a month  was associated with increased risk for HFMD. Therefore, hand-washing should be recommended to prevent HFMD, and the potential threat of airing out bedding should be carefully considered. However, further studies are needed to examine other possible risk factors. Enterovirus 71 (EV71), Coxsackievirus A16 (CA16), and Coxsackievirus A6 (CA6) [Hand, foot and mouth disease (HFMD) is a common acute infectious disease that is mainly caused by enteroviruses. In China, the major causative pathogens are A6 (CA6) . TransmiA6 (CA6) ,4.Outbreaks of HFMD occur worldwide, but these are more frequent in Asian countries . In 2007et al. [et al. [et al. [et al. [Until now, no effective chemoprophylaxis is available for HFMD. A vaccine for EV 71 has been recently developed, but has not been widely utilized . Moreoveet al.  mainly f [et al.  indicate [et al.  analyzed [et al.  exploredThis study is a non-matching case-control study focusing on HFMD cases among children 10 years old and younger. Cases were selected from Shijie Hospital of Dongguan City and Yue Bei People’s Hospital in Guangdong Province in between January 2015 and December 2015. We used the case definition of HFMD given in the “Guideline for the diagnosis and treatment of hand foot and mouth disease (2010 ed.)” issued by the Chinese Ministry of Health . HFMD isStandard questionnaire was used for data collection in face-to-face interviews administered by trained interviewers. Interviews were administered to the guardians of the subjects. The collected data comprised demographic factors , socio-economic factors , clinical factors , and behavior factors . All sets of information on risk factors were truncated to the date of onset for cases and to the date of the interview for the control subjects. Medical files were reviewed as needed for clarification. Considering the incubation period of HFMD, we probed the most exposures during a two week period leading to the onset of the disease for the cases or before the interview for the control subjects.t-test if the sets of data were normally distributed. Otherwise, the Kruskal-Wallis method was used. Univariate logistic regression analysis was first performed to examine the odds ratios (OR) for risk factors. Then age was included in the logistic regression analysis to adjust confounder effect. A further multivariate logistic regression analysis was then performed to examine the adjusted odds ratios for risk factors that were significant in previous multivariate analyses. A p value of <0.05 was considered statistically significant. Moreover, the receiver operating characteristic (ROC) curve was fitted to the observed data, and the accuracy of the model was measured by the area under the curve (AUC) [Database was built by EpiData3.1 , and statistical analysis was performed by SPSS19.0 . Descriptive analysis was performed for demographic information. Differences among the means of continuous variables were tested using Student’s ve (AUC) .p = 1). Overall, 126 male participants and 96 female participants were enrolled. Gender distributions were similar between the two groups. The average age (±SD) of participants is 4.47 (±2.07) years old, and participants infected with HFMD were younger than those in the control group. The age of cases were ranged from 1 to 8 years old, while the age of controls were ranged from 0 to 10 years old.Characteristics of both case and control participants were summarized in The results of univariate logistic regression are summarized in A multivariate logistic regression was performed to further investigate the association between HFMD and the four factors identified by the previous tests. Results are summarized in et al. [et al. [HFMD caused by enteroviruses continues to be a threat among Asian children . As no uet al.  and Xie  [et al. , both of [et al. . Hand-wa [et al. . Our stup value < 0.001). The results showed that the family has younger child tended to airing out their bedding more often. Furthermore, we found that airing out bedding is associated with an increased risk for HFMD and a dose-response effect was also observed. Airing out bedding under the sun is very common among Chinese residents to lower the humidity and dust mite levels of bedding. It is also recommended by some local Centers for Disease Control and Prevention as an intervention to prevent HFMD . HoweverWe also discovered that a recent history of cold food consumption may be related to protective effects, which were seldom mentioned in the previous studies. The cold food in our study refers to cold dish, sushi and ice-cream but not including fruits. This finding is inconsistent with a few current prevention guidelines in China . Since eOur findings have both academic value and practical significance. These results provide epidemiological evidence of the benefits of hand-washing and support the importance of hand-washing to prevent HFMD and other infectious diseases. This study also recognized the potential harm of airing out bedding, which was not mentioned in previous studies. In addition, with an AUC of 0.895, the final logistic model showed a good capacity of separating children with high risk of HFMD infection from the control group. For better conclusion, this study covered a more comprehensive range of possible risk factors, namely, demographic factors, birth and feeding conditions, and living habits of child and his/her caregivers. A few limitations of this study need to be considered. We selected the control subjects from the hospital information system and kindergartens randomly, instead of performing a matching case-control study, which may have led to a potential selection bias. To solve the problem of age distribution being different among the two groups, we adjusted the age in the analysis to eliminate the possible confounding effects. The study is also suffered from Berkson’s bias. The ORs in our results may be lower than the real value since some of the controls are unhealthy children with other diseases. These children may have poor immunity and high risk to get HFMD. Moreover, only HFMD patients who went to hospital were included as cases, which may not be representative for all HFMD infections. Recall bias should not be a serious problem in our study, as most exposures were determined retrospectively from the recall of caregivers. Interviews for the cases were performed when they went to the doctor, and the questions only covered the occurrences two weeks prior to the interview, which may have reduced the recall bias. Besides, as a cross-sectional study, we don’t have sufficient evidence to establish a causal relationship between risk factors and HFMD infection. Further randomized control trial or cohort study may be needed to provide stronger evidence. Another limitation is that we did not have a more detailed set of information describing the methods of hand-washing. The difference between the protective effects of hand-washing using water only and hand-washing with soap should be further explored. Socio-economic factors were not included as a risk factor, which is closely related with personal and environment hygiene. But we investigate whether being a permanent resident have any effect on HMFD infection, which may reflect the participant’s socio-economic status in some extent. This study suggests that hand-washing is a main strategy to prevent HFMD infection. Moreover, we need to be more careful when recommending airing out bedding as to the public. Further studies are warranted in the future because of potential limitations.


Conflicting renal effects of nesiritide have been reported in patients with acute decompensated heart failure. To answer this controversy, we performed a meta-analysis of randomized controlled trials to evaluate the influence of nesiritide on renal function in patients with acute decompensated heart failure.Articles were obtained from PubMed, Medline, Cochrane Library and reference review. Randomized controlled studies that investigated the effects of continuous infusion of nesiritide on renal function in adult patients with acute decompensated heart failure were included and analyzed. Fixed-effect model was used to estimate relative risk (RR) and weight mean difference (WMD). The quality assessment of each study, subgroup, sensitivity, and publication bias analyses were performed.P = 0.023). In subgroup analysis, high-dose nesiritide strongly associated with renal dysfunction , but no statistical differences were observed in standard-dose , low-dose groups  and same results were identified in the subgroup analysis of placebo controlled trials. Peak mean change of serum creatinine from baseline was no significant difference .Fifteen randomized controlled trials were eligible for inclusion. Most of included studies had relatively high quality and no publication bias was found. Overall, compared to control therapies, nesiritide might increase the risk of worsening renal function in patients with acute decompensated heart failure (RR 1.08, 95% CI 1.01–1.15, In our meta-analysis, nesiritide may have a dose-dependent effect on renal function in patients with acute decompensated heart failure. High-dose nesiritide is likely to increase the risk of worsening renal function, but standard-dose and low-dose nesiritide probably have no impact on renal function. These findings could be helpful to optimize the use of nesiritide in clinical practice. Worsening renal function (WRF) is prevalent in patients with acute decompensated heart failure (ADHF), and it may independently predict a worse outcome as well as increase the mortality of ADHF patients –3. HenceNesiritide, a recombinant B-type natriuretic peptide (BNP) with a potent vasodilatory property, has been approved for the treatment of ADHF, on the basis of its ability to reduce pulmonary-capillary wedge pressure and improve dyspnea , 5. AftePertinent articles were systematically searched  in PubMed, Medline and Cochrane Library by two trained investigators. The following search strategy developed according to Yan et al  was utilAn article was eligible for inclusion if it met the following criteria: 1) original articles; 2) RCT design; 3) adult (age ≥ 18 years) participants with ADHF; 4) nesiritide administered through intravenous infusion; 5) reporting the incidence of WRF or mean change of serum creatinine (SCr) from baseline. Articles which did not meet the aforementioned criteria were excluded from analysis. This meta-analysis was performed according to the Preferred Reporting Items for Meta-Analyses (PRISMA) statement checklist.Two authors independently extracted the pertinent data for the meta-analysis, with divergences resolved by discussion. The following information was abstracted from each study: characteristics of each study , characteristics of participants , intervention and control treatments  and clinical outcomes (incidence of WRF and peak mean change of SCr from baseline during the treatment of nesiritide). If several articles reported the same clinical trial, the one with the most complete data was included in analysis. And if a trial was designed with three or more than three groups, the control group was considered as a whole in the overall analysis and single group was considered separately in subgroup analysis.Two authors independently evaluated, with disagreement resolved by discussion, the risk of bias for the included RCTs in accordance with the following criteria developed by the Cochrane risk of bias tool: random sequence generation, allocation concealment, blinding of participants and personnel, blinding of outcome assessment, incomplete outcome data, selective reporting and other sources of bias.P ≤ 0.10 indicating significant heterogeneity) and I2 test. The relationship between nesiritide and incidence of WRF was analyzed using Mantel-haenszel (M-H) fixed-effect model (FFE) to calculate the relative risk (RR) and 95% confidence interval (CI), when there was no significant heterogeneity among the included studies; otherwise, a random-effect model was chosen. With regard to the analysis of peak mean change of SCr from baseline, inverse variance (IV) fixed-effect model was used to calculate the weight mean difference (WMD) and 95% CI without significant heterogeneity among the included studies; otherwise, a random-effect model was chosen. In order to evaluate the influence of different doses of nesiritide on the incidence of WRF, subgroup stratification analyses were also conducted. Sensitivity analysis was performed to evaluate the stability of results by exclusion of each study one by one. And publication bias was assessed with the use of funnel plots and Egger’s test.All the statistical analyses were performed with the use of STATA 12.0 . Heterogeneity was evaluated using chi-square test . The funnel plot demonstrated a little asymmetry , overall meta-analysis indicated that the use of nesiritide was associated with a significant increase in the risk of WRF . However, there was no significant influence of nesiritide on renal function in the standard-dose group including five RCTs or subgroups . Furthermore, in the low-dose group consisted of four RCTs or subgroups, significantly renal benefit was not observed either .To further investigate the relationship between continuous infusion doses of nesiritide and renal dysfunction, subgroup meta-analysis was performed. Five RCTs administered nesiritide in a high-dose manner. The data of analysis showed that high-dose nesiritide strongly increased the risk of renal dysfunction in patients with ADHF , but neither standard-dose  nor low-dose nesiritide  statistically increased the risk of WRF. Furthermore, to identify whether baseline chronic kidney disease (CKD) could influence the occurrence of WRF in patients with ADHF, a subgroup analysis based on the condition of baseline renal function was carried out. Interestingly, in comparison with control therapies, nesiritide significantly increased the risk of WRF in non-CKD patients  controlled trials to partly exclude the potential drug-drug interactions by removing the active controlled studies which used dobutamine, nitroglycerin or sodium nitroprusside as control treatments. As shown in P = 0.121; 2 = 22.3%, P = 0.252). Funnel plot did not show obvious asymmetry , standard-dose of nesiritide neither deteriorated nor improved the renal function of patients with ADHF. The conclusions of most RCTs included in these subset analyses were in accordance with our result , 17, 22.There are conflicting data about the effect of low-dose nesiritide on renal function. All the RCTs analyzed in our meta-analysis have not found that low-dose nesiritide had a potential to improve renal function in patients with ADHF , 18, 21.Data from our subgroup analysis indicated that nesiritide significantly increased the risk of WRF in non-CKD patients, however the adverse effect was not observed in ADHF patients with pre-existing CKD. This finding is a little confusing. But it was noteworthy that all the studies in CKD group administered nesiritide with either standard or low dose , 16, 22.SCr is still the main biomarker for predicting the renal function of patients . In our All the studies included in our meta-analysis were randomized design and had relatively high quality. Hence, our conclusions were relatively reliable, while there were several limitations in our study. Firstly, small sample size and limited number of clinical trials in subgroup analyses might produce bias. Secondly, the definition of WRF in included studies was not completely consistent, which might result in clinical and statistical heterogeneity. Thirdly, duration of intravenous nesiritide infusion was not same in each study, which might influence the clinical outcome. Fourthly, besides placebo (standard care), the control therapies also included dobutamine, nitroglycerin, and sodium nitroprusside. Therefore drug-drug interactions were not to be fully considered and might impact clinical outcome.In conclusion, the present meta-analysis demonstrates that nesiritide may have a dose-dependent effect on renal function in patients with ADHF. High-dose nesiritide may increase the risk of WRF, but neither standard-dose nor low-dose nesiritide is likely to have a significant impact on renal function in patients with ADHF. These findings could be helpful to optimize the use of nesiritide in clinical practice. In the future, more powered RCTs will be needed to confirm these findings. Additionally, whether low-dose nesiritide could improve the renal function in patients with ADHF also requires to be addressed.S1 PRISMA Checklist(DOC)Click here for additional data file.S1 FileFig A. Funnel plot of included studies for risk of WRF analysis. Fig B. Egger’s publication bias plot of included studies for risk of WRF analysis. Fig C. Funnel plot of included studies for peak mean change of SCr from baseline analysis. Fig D. Egger’s publication bias plot of included studies for peak mean change of SCr from baseline analysis.(DOCX)Click here for additional data file.S1 Table(DOCX)Click here for additional data file.


Moreover, the frequencies of CCR5-59029 with GG or GA genotype were higher than AA genotype in acute coronary syndrome individuals . In conclusion, we found that CCL5-403 polymorphism may increase genetic susceptibility of CAD. CCL5-403 or CCR5-59029 single nucleotide polymorphism may include genotype score and it may predict cardiovascular event.The chemokine receptor CCR5 polymorphism, which confers resistance to HIV infection, has been associated with reduced risk of cardiovascular disease. However, the association of the chemokine, CCL5, and its receptor, CCR5, polymorphism and coronary artery disease (CAD) in the Taiwanese has not been studied. In this study, 483 subjects who received elective coronary angiography were recruited from Chung Shan Medical University Hospital. CCL5-403 and CCR5-59029 were determined by polymerase chain reaction-restriction fragment length polymorphism. We found that CCL5-403 with TT genotype frequencies was significantly associated with the risk of CAD group (odds ratio = 3.063 and  The importance of coronary artery disease (CAD) in contemporary society is attested to by the almost epidemic number of persons . The WorChemokines (chemotactic cytokines) are small heparin-binding proteins that direct the movement of circulating leukocytes to sites of inflammation or atherosclerosis . The cheCCR5 was famous as a coreceptor for macrophage-tropic HIV-1 . A 32-baIn the recent decades, many serial studies about single nucleotide polymorphism of CCR5 and its ligand were published. However, the results of the different studies were diversity. CCR5 Δ32 had been identified as the reduced early onset coronary artery disease in women . In the CCL5 (RANTES) is one of the CCR5 ligands that have been linked to atherosclerosis. Simeoni et al. reported that RANTES A-403 was associated with CAD independently of conventional factors and hsCRP or fibrinogen as inflammatory biomarkers . The assIn the Asian study, Nakajima et al. found that the CCL5 promoter-28G genotype and CCR5 promoter 59029A genotype may be independent risk factors for diabetic nephropathy and may have an additive effect on nephropathy in the Japanese populations . HoweverThere were 483 Taiwanese adult patients with angina pectoris included in this study during the period from April 2007 through March 2009 in the Chung Shan Medical University Hospital. All of them received coronary angiography after the serial cardiovascular noninvasive examinations. The exclusive criteria were recent stroke, out-hospital cardiac arrest, and incomplete data. They were divided to CAD group  or non-CAD group  according to the result of coronary angiography. There were 348 male (108 non-CAD and 240 CAD) and 135 female (53 non-CAD and 82 CAD) patients in our study. All the patients were given an informed consent and were well told of the study protocol. Blood samples were collected via venous puncture and were analyzed by the central research laboratory. The study was approved by the Institutional Review Board of Chung Shan Medical University Hospital (CSMUH number CS07095), and informed consent was obtained from each participant.Coronary angiography was performed via transradial or transfemoral approach. The lumen stenosis of coronary artery was measured by quantitative coronary angiography (QCA). Angiographic criteria defining CAD cases or controls were as follows: (1) cases group: lumen stenosis > 50% lumen narrowing on one or more major epicardial coronary arteries, (2) control group: lumen stenosis < 50% lumen narrowing on one or more major epicardial coronary artery segments.μL volume containing 100 ng DNA template, 1.0 μL of 10x PCR buffer , 0.25 U of Taq DNA polymerase (Invitrogen), 0.2 mM dNTPs , and 200 nM of each primer . The PCR cycling conditions were performed: 94°C for 5 min, followed by 35 cycles of 1 min at 94°C, 1 min at 57°C, and 2 min at 72°C, with a final step at 72°C for 20 min to allow a complete extension of all PCR fragments. For CCL5-403, the wild homozygous alleles (C/C) yielded a 206-bp product, and the heterozygous alleles (C/T) yielded 206-, 180-, and 26-bp products, while the mutant homozygous alleles (T/T) yielded 180- and 26-bp products. For CCR5-59029, the wild homozygous alleles (G/G) yielded 134-bp products, and the heterozygous alleles (G/A) yielded 268- and 134-bp products, while the mutant homozygous alleles (A/A) yielded a 268-bp product . Genotyping for CCL5-403 and its receptor CCR5-59029 was determined by PCR-restriction fragment length polymorphism assay (PCR-RFLP). Sequences of primers used for analysis of CCL5-403 rs2107538 C/T genotype were 5′-CACAAGAGGACTCATTCCAACTCA-3′ (forward primer) and 5′-GTTCCTGCTTATTCATTACAGATCGTA-3′ (reverse primer) to yield a product of 206 bps. The CCR5-59029 rs1799987 G/A polymorphism was amplified with the following primer: 5′-CCCGTGAGCCCATAGTTAAAACTC-3′ (forward primer) and 5′-TCACAGGGCTTTTCAA CAGTAAGG-3′ (reverse primer) (268 bp). We used the endonucleases Rsa I and Bsp 1286I for restricted enzyme. Polymerase chain reaction was performed in a 10  product . MoreoveU test and Fisher's exact test were used to compare the differences of age as well as demographic characteristics distributions between control group and cases group with CAD, since the small sample size was present in some categorical variables. The odds ratios (ORs) with their 95% confidence intervals (CIs) of the association between genotype frequencies and CAD susceptibility as well as clinical characteristics were estimated by multiple logistic regression models. A p value < 0.05 was considered significant. The data were analyzed on SAS statistical software . We assessed Hardy-Weinberg equilibrium by using a goodness-of-fit χ2 test for biallelic markers.The average age is presented as the mean ± SD. A Mann-Whitney n = 322) and control group (n = 161). In the two groups, there were no significant differences in gender, age, height, weight, smoking, family history, hyperlipidemia, and diabetes. However, significant differences were observed between case group and control group for characteristics like hypertension, atrial fibrillation, high CAD risk factors, acute coronary syndrome, aspirin used, heart failure, and recent angina (p < 0.05).The clinical characteristics of this study were summarized in p = 0.012). TT homozygote of CCL5-403 was 3.063-fold significant increasing risk in CAD group compared to CC or CT individuals . T allele frequency of CCL5-403 was also significantly different between cases and controls . However, GA or AA genotype frequency of CCR5-59029 did not show any statistically significant difference . An allele frequency of CCR5-59029 was not found significantly different among the cases group and control group .The genotype frequencies and distribution in cases and control groups were shown in We further analyzed the association between CCL5-403, CCR5-59029 genotype, and the demographics features in p = 0.008).We also further analyzed the association between CCL5-403, CCR5-59029 genotype, and the clinical characteristics in Tables α-granules together with PF-4/CXCL4 and deposited on the inflamed endothelium [The chemokine receptor CCR5 was well known as the coreceptor for HIV infection of macrophages. CCR5 deletion had reported the resistance to HIV infection and was found to reduce cardiovascular risk in animal models . CCL5 acothelium . CCL5 anothelium . Gene pop = 0.012). T allele odds ratio of CCL5-403 was 1.377-fold significant increasing risk in CAD group compared to C allele . The results were compatible with the findings of Simeoni et al. [p = 0.008). The association between CCR5-59029 and acute coronary syndrome was not shown in other reports. GA or AA genotype frequency of CCR5-59029 did not show any statistically significant difference. Thus, we assumed CCR5-59029 polymorphism was associated with acute coronary syndrome but not chronic stable coronary artery disease. Additionally, knock-down CCL5 expression in vascular muscle was found to reduce neointimal thickening and macrophage infiltration in mouse model [In our case control study, TT homozygote of CCL5-403 was 3.063-fold significant increasing risk in CAD group compared to CC or CT individuals (95% CI = 1.231–7.624, i et al. . They foi et al.  and Indii et al. . Most imse model . In summThere were many chemokines and their receptors play a role in the development of atherosclerosis. Specially, evidence was established as the important roles and key plays for CCL2/CCR2, CX3CL1/CX3CR1, and CCL5/CCR5 . The polThere were numerous studies investigating whether serum CCL5 levels were associated with cardiovascular disease, although it should be noted that levels of circulating chemokine do not correspond to levels in tissue. In Kraaijeveld et al.'s study, high circulating levels of CCL5 might be a marker for unstable angina . In contThe limitations in this study are shown below. Because we divided all patients to cases group and control group by the result of coronary angiography, allocation bias may be found in this study. For example, the percentage of atrial fibrillation in cases group is only 9.3% and the result is very lower than the 24–46% prevalence of atrial fibrillation in the previous report . 24 hourIn conclusion, CCL5-403 polymorphism may increase genetic susceptibility of CAD. CCL5-403 with TT genotype or T allele is associated with CAD. CCR5-59029 polymorphism is associated with acute coronary syndrome but not chronic stable coronary artery disease. The SNPs of CCL5/CCR5 can be the predictive chemokine factor except for CCL2/CCR2 or CX3CL1/CX3CR1 polymorphism in the Asian population.


UCP) 1–3 polymorphisms and susceptibility to obesity has been investigated in several genetic studies. However, the impact of these polymorphisms on obesity is still under debate, with contradictory results being reported. Until this date, no meta-analysis evaluated the association of UCP polymorphisms with body mass index (BMI) variability. Thus, this paper describe a meta-analysis conducted to evaluate if the -3826A/G (UCP1); -866G/A, Ala55Val and Ins/Del (UCP2) and -55C/T (UCP3) polymorphisms are associated with BMI changes.The relationship between uncoupling protein (UCP1-3 polymorphisms and BMI. Weighted mean differences (WMD) were calculated for different inheritance models.A literature search was run to identify all studies that investigated associations between UCP2 55Val/Val genotype was associated with increased BMI in Europeans . Moreover, the UCP2 Ins allele and UCP3-55T/T genotype were associated with increased BMI in Asians . However, a decreased BMI mean was observed for the UCP2-866 A allele in Europeans under a dominant model of inheritance . There was no significant association of the UCP1-3826A/G polymorphism with BMI mean differences.Fifty-six studies were eligible for inclusion in the meta-analysis. Meta-analysis results showed that UCP2-866G/A, Ins/Del, Ala55Val and UCP3-55C/T polymorphisms and BMI mean differences.The meta-analysis detected a significant association between the  Obesity is the most common nutritional disease in industrialized countries, constituting a priority health problem. It is associated with the development of cardiovascular diseases, type 2 diabetes mellitus and various cancers, which leads to higher morbidity and mortality rates, reducing the quality and expectancy of life of sufferers Uncoupling proteins 1, 2 and 3 are members of an anion-carrier protein family located in the mitochondrial inner membrane UCP1-3 loci and susceptibility to obesity has been investigated in several genetic studies and particular attention has been focused on the -3826A/G (rs1800592) polymorphism in the UCP1 gene, the -866G/A polymorphism (rs659366), the Ala55Val  polymorphism and the Ins/Del polymorphism in the UCP2 gene, and the -55C/T (rs1800849) polymorphism in the UCP3 gene et al. UCP2-886G/A, UCP2 Ala55Val, and UCP3-55C/T polymorphisms with obesity in a meta-analysis of 22 articles. The meta-analysis revealed that the UCP2-866G/A polymorphism may be a risk factor for obesity in Europeans, but not in Asian subjects. In agreement with this data, Andersen et al. UCP2-866G/A polymorphism and obesity in subjects of European descent. Until this date, no meta-analysis evaluated the association of UCP polymorphisms with body mass index (BMI) variability. Thus, to further investigate the potential role of UCP1-3 polymorphisms in influencing BMI changes, we performed a systematic review and meta-analysis of the literature on the subject.The relationship between UCP1-3 genes: -3826A/G (rs1800592) polymorphism in the promoter region of the UCP1 gene, -866G/A (rs659366) polymorphism in the promoter region, the Ala55Val  polymorphism in exon 4 and the Ins/Del polymorphism, which is an insertion/deletion of 45 bp in the 3′ untranslated region (3′UTR) of exon 8 of the UCP2 gene, and the -55C/T (rs1800849) polymorphism in the promoter region of the UCP3 gene. The following medical subject headings (MeSH) were searched: (‘‘Obesity’’ OR “Body mass index”) AND  AND . The search was limited to human studies and English or Spanish language papers and was completed on January, 2014. All of the articles identified were also searched manually to identify any other relevant citations.This study was designed and described in accordance with current guidelines for execution of systematic reviews and meta-analyses UCP1-3 polymorphisms in question. Studies were excluded from the analysis if they did not have sufficient data or if they did not employ validated genotyping methods. If data were duplicated and had been published more than once, the most complete study was chosen.Two investigators (B.M.S and A.P.B.) independently reviewed the titles and abstracts of all articles selected in order to evaluate whether the studies were eligible for inclusion in the meta-analysis. Disagreements were resolved by discussion between them and when necessary a third reviewer (D.C.) was consulted. When abstracts did not provide sufficient information to fulfill the inclusion and exclusion criteria, the full text of the article was retrieved for evaluation. We included observational studies that compared BMI (mean ± SD) among different genotypes of one or more of the UCP1-3 polymorphisms.Data were independently extracted by two investigators (L.A.B. and T.S.A.) using a standardized abstraction form, and consensus was sought in all extracted items. When consensus could not be reached, differences in data extraction were resolved by referencing the original publication or by consulting a third reviewer (D.C.). The information extracted from each individual study was as follows: name of first author, publication year, ethnicity, number of subjects, age, gender, and BMI in different genotypes of 2 test. Some articles only reported the mean BMI according to the different genotypes after stratification by sex or other characteristic such as presence of diabetes or obesity, so we had to calculate the pooled mean BMI to include this information in our meta-analysis. Heterogeneity among studies was tested using a χ2-based Cochran’s Q statistic and inconsistency was assessed with the I2 metric Q statistic and I2>50% for the I2 metric statistics. Where significant heterogeneity was detected, the DerSimonian and Laird random effect model (REM) was used to calculate weighted mean differences (95% CI) in BMI for each individual study and for the pooled effect; where heterogeneity was not significant, the fixed effect model (FEM) was used for this calculation Subjects’ genotype distributions were tested for conformity with Hardy-Weinberg Equilibrium (HWE) using a goodness-of-fitness χUCP polymorphisms show variable frequencies across ethnic groups. Risk of publication bias was assessed using funnel plot graphics, analyzed both visually and with the Begg and Egger test Meta-regression and sensitivity analyses were carried out to identify key studies with a substantial impact on inter-study heterogeneity. The factors investigated by meta-regression were age, gender and ethnicity. Sensitivity analyses were performed after stratifying the studies by ethnicity, given that the UCP polymorphisms, but not the ones of interest here. A total of 56 articles fulfilled the eligibility criteria and were included in the meta-analyses Table S1. The median age of analyzed subjects was 46.0  years, and the median BMI was 37.0 (20.3–50.2). Sixty-one percent of the subjects were men, 25.3% had diabetes, and 57.1% had obesity. Most of the studies were performed in European populations.Table S2 shows the mean BMI ± SD according to the different UCP1-3 genotypes in all studies included in the meta-analysis. Twenty studies analyzed the UCP1-3826A/G polymorphism (6191 subjects), 18 studies analyzed the UCP2-886G/A polymorphism (12072 subjects), 6 analyzed the UCP2 Ala55Val polymorphism (15060 subjects), 8 analyzed the UCP2 Ins/Del polymorphism (8490 subjects), and 8 investigated the UCP3-55C/T polymorphism (6663 subjects).UCP polymorphisms and differences in mean BMI values. UCP1-3826A/G and UCP3-55C/T polymorphisms in Europeans and Asians, respectively, assuming a dominant inheritance model. UCP2-866G/A, Ala55Val and Ins/Del polymorphisms in Europeans and Asians, respectively, also under a dominant model.UCP1-3826A/G and UCP3-55C/T polymorphisms were not associated with any change in BMI irrespectively of the inheritance model or after stratification by ethnicity  but not in Asians. On the other hand, the Ins allele of the UCP2 Ins/Del polymorphism was associated with increased WMD in BMI in Asians  but not in Europeans. The UCP2 Ala55Val polymorphism was also associated with increased WMD in BMI in Europeans, under a recessive model .2>50%) among studies in almost all comparisons of BMI according to the UCP1-3 genotypes under the different inheritance models. To investigate this finding in greater depth, sex, age and ethnicity were used as covariates in meta-regression analyses performed for the five UCP polymorphisms under different inheritance models. None of the covariates used in univariate and multivariate meta-regression analyses could individually explain the heterogeneities observed (data not shown).As can be observed in UCP2 Ala55Val polymorphism (data not shown). In contrast, some studies seem to be responsible for the inter-study heterogeneity observed in the meta-analyses of UCP1-3826A/G, UCP2-866G/A and Ins/Del, and UCP3-55C/T polymorphisms  in Europeans. However, this polymorphism causes a conservative amino acid change at position 55 of exon 4 and, until now, there had been no evidence that this alteration generates a functional change in the protein UCP2/UCP3 loci. The biological significance of the UCP2 Ins/Del polymorphism is not well known. However, its location in the 3′UTR suggests that this polymorphism could alter the mRNA processing or the stability of the transcript. Any reduction in the UCP2 mRNA stability could compromise the ability to remove excess calories through thermogenesis, especially in a person with a propensity to obesity from other genetic or environmental influences UCP2 Ins/Del polymorphism is associated with increased BMI in Asians.Our data indicate that the UCP3-55C/T polymorphism is located 55 bp upstream of the most commonly used transcription initiation site of skeletal muscle UCP3-55C/T polymorphism on UCP3 expression as well as to confirm its association with obesity. It is worth noting that a previous meta-analysis from our group showed an association of the UCP3-55C/T polymorphism with risk for type 2 diabetes mellitus in Asians  The UCP polymorphisms analyzed. To investigate this problem in greater depth, meta-regression analyses were performed and showed that age, gender and ethnicity did not make significant contributions to inter-studies heterogeneity. The heterogeneity observed could be due to differences in sample selection, genotyping methods or gene-environment interactions and, without detailed information on the metabolic and clinical characteristics of the studies reviewed; we could not fully exclude the possibility that the heterogeneity observed might reduce our power to detect true associations. “Leave one out” sensitivity analyses showed that after excluding some studies, the heterogeneity for these analyses decreased significantly. However, despite the exclusion of these studies, only the pooled WMD obtained for the UCP3-55C/T polymorphism changed significantly: after the exclusion of Hamada et al. UCP3-55C/T polymorphism became associated with increase BMI in Asians. Fourth, obesity is a complex disorder that involves interactions of genes and environment. Many studies revealed different effects of the UCP1-3 polymorphisms depending on the physical activity and lifestyles UCP polymorphisms and BMI changes after stratifying by ethnicity. For the whole sample, we had at least 80% power (α = 0.05) to detect even modest changes in WMD for all analyzed polymorphisms, which is an evidence that our results are robust.The results of the present meta-analysis should be interpreted within the context of a few limitations. First, meta-analysis is prone to publication bias, and although we have attempted to trace unpublished studies, we cannot be sure that small negative studies were overlooked. Moreover, we did not detect any significant publication bias in our analysis, showing that our data is statistically reliable. Second, because of the difficulty in getting the full texts of articles published in several languages, we only included studies published in English and Spanish. Third, inter-studies heterogeneity is common in meta-analyses for genetic association studies UCP1-3826A/G polymorphism is not associated with significant changes in BMI. However, our results suggest that the UCP2-866A allele is associated with decreased BMI in Europeans, while the UCP2 Val55Val genotype is associated with increased BMI in Europeans. The UCP2 Ins allele and the UCP3-55T/T genotype seem to be associated with increased BMI in Asians. It is important to keep in mind that BMI is one phenotype of obesity; thus, we can not exclude the possibility that these polymorphisms could have different effects in other phenotypes, such as fat body mass index, waist circumference, and intra-abdominal adipose tissue. Since small sample sizes were obtained for some of the analyses performed in Asians, further additional larger studies that allow stratification for ethnicity and gene-gene and gene-environment interactions should also be conducted in order to elucidate the roles possibly played by UCP polymorphisms in obesity and BMI changes.In conclusion, our results indicate that the Figure S1Funnel plot for studies of the UCP1-3826A/G polymorphism under a dominant model of inheritance.(TIF)Click here for additional data file.Figure S2Funnel plot for studies of the UCP2-866G/A, Ala55Val and Ins/Del polymorphisms under a dominant model of inheritance.(TIF)Click here for additional data file.Figure S3Funnel plot for studies of the UCP3-55C/T polymorphism under a dominant model of inheritance.(TIF)Click here for additional data file.Table S1Characteristics of the studies included in the meta-analyses.(XLS)Click here for additional data file.Table S2Body mass index means according to different genotypes of UCP1-3826A/G, UCP2-886G/A, UCP2 Ala55Val, UCP2 Ins/del and UCP3-55C/T polymorphisms for the studies included in the meta-analysis.(DOC)Click here for additional data file.Checklist S1PRISMA Checklist.(DOC)Click here for additional data file.Flow Diagram S1PRISMA Flow Diagram.(DOC)Click here for additional data file.


Catalase (CAT), one antioxidant enzyme, may provide resistance against many diseases. Many previous studies reported predictive and prognostic values of CAT C262T polymorphism in cancers, with divergent results. This study aimed to summarize the overall relationships between CAT C262T polymorphism and cancer risk or survival. A total of 27 eligible publications were included in susceptibility analysis, while 8 publications contained survival outcomes. The results revealed significant relationship between CAT C262T polymorphism and cancer risk, subgroup analyses indicated the CAT C262T polymorphism was significantly correlated with an increased risk for prostate cancer  and increased risk among Caucasians , while no associations between the polymorphism and Asian or mixed population were established. In the survival analysis, no interactions were identified between this polymorphism and cancer survival . In conclusion, the CAT C262T polymorphismmay be a candidate markerfor cancer risk with type-specific and population-specific effects but not a fine prognostic factor for cancer survival. The molecular mechanisms of carcinogenesis have not been wellunderstood, but growing studies have reported that oxidative stress played a significant role in the progression of many diseases, including cancers3467Catalase is an important endogenous antioxidant enzyme thatcatalyzes hydrogen peroxide into oxygen and water, thus neutralizing the deleterious effects of ROS11616The initial search yielded 1676 articlesthrough the databases of Pubmed, Embase and China National Knowledge Infrastructure (CNKI). After screening the titles and abstracts, 82 potentially relevant articles were retrieved for the full-text. 49 articles were excluded: 3 were reviews; 9 were conference abstracts; 4 were related to other SNPs of the CAT gene; 11 did not report extractable data; 22 were irrelevant papers. Finally, a total of 33 articles678111215161718192021222324252627282930313233343536373839404167811121516171819212324252627282930313334353738396202229323641The meta-analysis of the 27 articles678111215161718192123242526272829303133343537383967161825The meta-analysis included 8 studies investigating CAT C262T polymorphism and cancer survival6202229323641We didn’t detect any significant publication bias by Begg’ test . The publication language and publication date were not restricted in our search. Some potential publications were obtained from a manual search of the references of retrieved articles.The inclusion criteria were: (1) case-control studies or cohort studies; (2) evaluating the associations between the CAT C262T polymorphism and cancer risk or survival outcomes; (3) detailed data on genotype frequency of the CAT C262T for calculating the odds ratios (ORs), available hazards ratios (HRs) and 95% confidence intervals (95%CIs). The exclusion criteria were: (1) reviews, conference abstracts, case reports, animal studies or editorials; (2)without available genotype frequency of the CAT C262T; (3) when the same or overlapped population and duplicated studies were met, only the most recent studies with sufficient information were included.Two investigators extracted data independently and consensus on all the items was reached after discussion. The main information included the first author, publication year, country, ethnicity, source of controls, sample, quality control, quality health, cancer type, number of cases and controls, genotype distributions of cases and controls, genotyping method, HWE of the control groups, and HR with 95%CI of this polymorphism in survival analysis.2 statistics. If I2 > 50%, the random-effects model was applied to analysis; otherwise, the fixed-effects model was adopted48All statistical analyses were conducted with STATA 12.0 . The statistical heterogeneity among the studies was calculated by the IHow to cite this article: Wang, C.-D. et al. The Role of Catalase C-262T Gene Polymorphism in the Susceptibility and Survival of Cancers. Sci. Rep.6, 26973; doi: 10.1038/srep26973 (2016).


The association between long-term benzodiazepine use and risk of dementia remains controversial. Therefore, current study aimed to quantify this association, and to explore a potential dose–response pattern.We searched PubMed, Embase and the Cochrane Library through August 17, 2014. We included nested case-control or prospective cohort studies that provided risk estimates on the association of benzodiazepine use with risk of dementia, and a clear definition of status of benzodiazepine use. Overall effect size was calculated using a random-effects model.Six studies were eligible for inclusion, involving 11,891 dementia cases and 45,391 participants. Compared with never users, pooled adjusted risk ratios (RRs) for dementia were 1.49  1.30–1.72) for ever users, 1.55 (95% CI 1.31–1.83) for recent users, and 1.55 (95% CI 1.17–2.03) for past users. The risk of dementia increased by 22% for every additional 20 defined daily dose per year . When we restricted our meta-analyses to unadjusted RRs, all initial significant associations persisted.Long-term benzodiazepine users have an increased risk of dementia compared with never users. However, findings from our study should be treated with caution due to limited studies and potential reverse causation. Large prospective cohort studies with long follow-up duration are warranted to confirm these findings. As the number of older adults grows dramatically worldwide, dementia is regarded as a public health priority that has imposed an increased burden on our society in either financial or psychological respect. Nearly 35.6 million people were afflicted with dementia in 2012, and this number is expected to double by 2030 .The guidelines recommend that the overall duration of benzodiazepine use should be limited to several weeks , but lonMeta-analysis is usually described as a quantitative statistical method that improves statistical power through pooling effect sizes from individual studies, and plays an important role in solving controversial issues in medicine. Therefore, we conducted a meta-analysis to quantify the relationship between long-term benzodiazepine use and dementia. Moreover, we explored a potential dose–response pattern between long-term benzodiazepine use and dementia.We performed this study following the PRISMA statement. Two researchers (G.C.Z. and Y.W.) independently performed an electronic search of PubMed, Embase and the Cochrane Library through August 17, 2014, with the following keywords: “dementia”, “Alzheimer's disease”, “Alzheimer disease”, “vascular dementia”, “benzodiazepines”, “benzodiazepine”, “hypnotics”, “hypnotic”, “sedatives” and “sedative”. We limited our searches to human subjects. The reference lists of identified studies and pertinent reviews were checked for identifying additional citations. No effort was attempted to obtain extra information by contacting the authors of included studies via e-mails.Nested case-control or prospective cohort studies that provided a clear definition of status of benzodiazepine use  and adjusted risk estimates on the association between benzodiazepine use and risk of dementia met our inclusion criteria. We excluded studies that compared recent users with a reference group that indiscriminately pooled both past and never users.Two reviewers (G.C.Z. and Y.W.) independently performed study screening with a two-stage method. In the first stage, we read titles and abstracts to exclude obviously irrelevant studies. In the second stage, we read the full text to further exclude unrelated studies. Any discrepancy about eligibility of studies was resolved by discussion.Two reviewers (G.C.Z. and Y.W.) independently extracted following information: last name of the first author, publication date, study location, mean age of participants at baseline, number of cases and participants, maximum follow-up duration, exposure assessment, outcome assessment, status of benzodiazepine use and corresponding definitions, diagnostic criteria, the crude risk estimates with 95% confidence interval (CI), the most fully adjusted risk estimates with 95% CI, and adjustment factors. Any disagreement was settled by discussion. When multiple reports derived from the same cohort, we extracted data from the report with the longest follow-up duration.p<0.10). The I2 statistic, representing the percentage of the variability in risk estimates that is caused by heterogeneity rather than chance [2 <50%, moderate heterogeneity if 50%≦I2≦75%, and substantial heterogeneity if I2>75%. For two studies [We used risk ratio (RR) to assess the association between benzodiazepine use and risk of dementia. Hazard ratio and odds ratio were directly regarded as RR because dementia is a relatively uncommon event , 19. Then chance , was empn chance . Summary studies , 16 thatWe explored a dose–response pattern between benzodiazepine use and dementia on the basis of the method previously proposed by Orsini and colleagues . The indp<0.05 under two-sided test unless otherwise specified.We employed the Newcastle–Ottawa quality assessment scale  to judgeA sum of 1573 records  were identified through our initial database search. After removing duplicates, 984 relevant records remained. Of these remained records, 978 were judged to be not eligible for the inclusion using our two-stage method. Thus, 6 studies were included in our review and analysis .The characteristics of included studies are presented in p = 0.19; I2 = 35.1%).Although six included studies all provided risk estimates associated with ever use of benzodiazepines, two of them , 27 derip = 0.32; I2 = 15.0%).Current meta-analysis for recent use of benzodiazepine and risk of dementia included five studies , 16, 27,p<0.01; I2 = 72.6%).Our review included five studies , 16, 27 2 = 0.0%).The results of sensitivity analyses are shown in p = 0.32; I2 = 0.0%).On the basis of only two studies , 15, we p>0.05).We did not find the evidence of publication bias for any association through Begg’s test and Egger’s test  found that the long-term users of benzodiazepines have a 1.5- to 2-fold increased risk of developing dementia compared with never users . On the The prodromal symptoms of dementia, including sleep disturbance, anxiety and depression –31, can The prevalence of psychotropic medication use among the elderly is high , 35, witOf included studies, ever use of benzodiazepines was consistently defined as benzodiazepine use during follow-up, namely use of benzodiazepines at least once before the index date. The consistent definition increased the reliability of our pooled results for the association between ever use of benzodiazepines and dementia. However, we observed that the cut-off points between recent use and past use among included studies varied from 2 years  to 12 yep values for the difference between two conditions calculated from meta-regression (data not shown), we found that there was no significant difference in magnitude of the pooled effect size between two conditions. Therefore, pooled results from unadjusted and adjusted risk estimates resulted in virtually identical interpretation for the association of benzodiazepine use with risk of dementia. Consequently, calculation of unadjusted and adjusted summary RRs makes our findings more convincing rather than more obscuring.In this meta-analysis, we calculated summary RRs from unadjusted and adjusted data. Interestingly, a significantly positive association of benzodiazepine use with risk of dementia was consistently observed in both conditions. Furthermore, on the basis of In the current review, we found the evidence of moderate heterogeneity for pooled adjusted RRs on past use of benzodiazepines and dementia. As revealed by our sensitivity analyses, one study  that conOur review has several limitations. First, inclusion of a small number of studies did not permit us to investigate the potential effect modifiers for the association of benzodiazepine use and dementia through subgroup analyses. For example, Alzheimer’s disease and vascular dementia are the main subtypes of dementia, and they have different incidence and etiology. However, we could not confirm whether the risk of Alzheimer’s disease is different from that of vascular dementia due to limited numbers of included studies. Second, although there was no evidence of publication bias, we could not completely rule it out because Begg’s test and Egger’s test have limited statistical power when there are relatively small included studies. Third, we observed moderate heterogeneity for pooled results on past use of benzodiazepines and dementia. Such heterogeneity raised concerns about the reliability of pooled results. Nevertheless, we were capable of identifying the source of heterogeneity through our sensitivity analyses. Finally, our pooled results might be subject to the uncontrolled or residual confounding, although we used the most fully adjusted risk estimates. Nevertheless, when we pooled unadjusted RRs, the increased risk of dementia in benzodiazepine users remained significant.Prevention of dementia is critical under the condition that the pharmaceutical therapies for dementia fail to function well against disease progression . Long-teOn the basis of either unadjusted or adjusted risk estimates, our study consistently indicates that long-term benzodiazepine use is associated with an increased risk of dementia. Due to limited studies, especially dose-response analysis, and potential reverse causation, these findings should be treated with caution. Large prospective cohort studies with long follow-up duration are needed to confirm whether the association between long-term benzodiazepine use and increased risk of dementia is causal. If confirmed, long-term benzodiazepine use should be considered as a critical public health issue in the context of the widespread use of benzodiazepines and the huge burden of dementia across many countries.S1 Checklist(DOC)Click here for additional data file.S1 Dataset(XLS)Click here for additional data file.


ACE) insertion/deletion (I/D) polymorphism have long been linked to sporadic Alzheimer disease (SAD), but the established data remained controversial. To clarify this inconsistency, a comprehensive meta-analysis was conducted. Through searching of Pubmed, Embase, Alzgene, China National Knowledge Infrastructure (CNKI) and manually searching relevant references, 53 independent studies from 48 articles were included, involving a total of 8153 cases and 14932 controls. The strength of association was assessed by using odds ratios (ORs) with 95% confidence intervals (CIs). Further stratified analyses and heterogeneity analyses were tested, as was publication bias. Overall, significant associations were revealed between I/D polymorphism and SAD risk using allelic comparison , homozygote comparison  and the dominant model , but they were not sufficiently robust to withstand the false-positive report probability (FPRP) analyses. Otherwise, in subgroup analyses restricted to the high quality studies, the large sample size studies and studies with population-based controls, no significant association was observed in any genetic models. In summary, the current meta-analysis suggested that the ACE I/D polymorphism is unlikely to be a major determining factor in the development of SAD.Angiotensin-converting enzyme gene ( Alzheimer's disease (AD) is the most common form of dementia in the elderly, and it is characterized by progressive memory loss and cognitive dysfunction ACE), a zinc metalloprotease widely expressed in brain, as a possible modulator of Aβ metabolism ACE gene is located on chromosome 17q23 and consists of 26 exons and 25 introns. The most common polymorphism of ACE gene is the insertion/deletion (I/D) variant of 287 base pairs in intron 16, which has been suggested to be associated with serum ACE protein levels ACE protein domain ACE gene promoter region Recently, growing evidence has implicated angiotensin-converting enzyme (ACE) and (polymorphism or variant or genotype)”. In addition, the reference lists of reviews and retrieved articles were checked for potential studies. Articles that reported results from more than 1 population were considered as separate studies. Only studies published in English or Chinese were included.This meta-analysis was performed according to the methodology advocated by the PRISMA statement The studies included in the meta-analysis were required to meet the following criteria: (1) case-control or cohort design; (2) association between I/D polymorphism and SAD risk; (3) application of standardized clinical or pathologic criteria for the diagnosis of SAD; (4) sufficient genotype distributions for calculation of odds ratios (ORs) with 95% confidence intervals (CIs);ACE polymorphisms. When the articles contained duplicated data, the most recent or complete studies were selected.The following were excluded: (1) reports with incomplete data; (2) review articles, abstracts, case reports; (3) studies based on familiar Alzheimer's disease (FAD) or mild cognitive impairment (MCI); (4) studies about other Data were extracted from eligible articles independently by two of the authors, with any disagreement resolved by consensus. The following information was collected in a predefined data collection form: first author's name, publication year, country, geographical location of participants , sample size, AD diagnosis criteria, genotyping methods, source of controls, risk allele frequency in controls, results of Hardy-Weinberg equilibrium (HWE) in controls and quality assessment of studies.The quality of each study was independently assessed by two authors who used quality scoring criteria modified from previous studies . Second, genotype distributions of controls were used to estimate the frequency of the putative risk allele  in various geographic location using the inverse variance method First, deviance from HWE was assessed for the controls of each study using the chi-squared testQ statistic was used to test for heterogeneity, and the percentage variability of the heterogeneity between studies was quantified using the 2I statistic. Thus, 2I values around 25%, 50% and 75% would indicate low, medium and high heterogeneity respectively . The random-effect model (DerSimonian-Laird) 2I (%)>50% or P (Q)<0.10. Otherwise, the fixed-effect model (Mantel-Haenszel) Cochran's For each statistically significant association, the false-positive report probability (FPRP) analyses were performed using the method reported by Wacholder et al ACE SNPs; two articles were excluded because of FAD A total of 166 relevant articles were initially identified from Pubmed, EMBASE, Alzgene and CNKI databases. After titles and abstracts were screened, 93 articles were excluded because of irrelevant data. The full texts of the remaining 73 records were carefully reviewed . Among tACE I allele in control populations demonstrated variation among geographic location/ethnicity groups. Frequencies of the I allele were highest among Asians , followed by North Europeans  and South Caucasians . Sensitivity analyses excluding the HWE-deviated studies showed similar results.Overall, the eligible studies included 8153 cases and 14932 controls, and all these samples were genotyped. The pooled frequencies of the Summaries of the odds ratios for different comparisons were provided in When studies were stratified by sample size  and qualWe also investigated potential influence arising from the use of different genotyping method  , or in sample selection , or in methods , or it may be due to interaction with other risk factors . The meta-regression analyses indicated that the potential sources of heterogeneity for heterozygote comparison were sample size and publication time. However, the sources of heterogeneity for the other models were not found, suggesting that heterogeneity might also be explained by other confounding factors. Nevertheless, when studies were stratified by geographic location, the heterogeneity was higher in the Asian subgroup while it was decreased in other populations. It was the same with the results of Galbraith plot analyses, which spotted at least 14 studies as the outliers and 9 of these 14 studies belonged to the Asian subgroup. These two analyses provided evidence that a combination of heterogeneous studies from the Asian subgroup contributed to the moderate heterogeneity of overall analyses. However, in the Asian subgroup, meta-regression did not find any sources of heterogeneity (data not shown), suggesting that the heterogeneity in the Asian subgroup might be explained by other confounding factors. In general, more rigorous and uniform studies were required.ACE I allele frequency of the controls showed a modest difference across ethnicities . These inconsistent data may be explained by the different genetic background across ethnicities. Nevertheless, owing to the greater FPRP values of significant associations in the North European subgroup, the observed ethnic difference in this meta-analysis was also likely to be caused by chance because continued reliance on the standard p value criterion of 0.05 to define statistical significance without consideration of power or prior probability may generated a fluctuated risk estimate In the present study, results from populations with different genetic backgrounds were not the same. The combinations of the South Caucasians studies and the Asian studies showed no significant results. However, results from the North European subgroup were distinct and the pooled When studies were stratified by sample size, significant elevated SAD risks were associated with the I/D polymorphism in the small sample size subgroup. However, small sample with limited participants was often accompanied with selection biases, and lacked sufficient power to support or deny an association In the stratification analysis by source of controls, significant associations were observed using heterozygote comparison and the dominant model in the non populations-based subgroup. However, the genotype distributions in the non population-based studies may not be representative of the general population. Given the fact that these associations were not noteworthy  and no significant association was found in the population-based studies, we thought that this significant association for the I/D variant might be a spurious finding. More and larger population-based studies were required to further clarify the association between I/D variant and SAD risk.APOE ε4 status, gender, etc.Despite our efforts in performing a deeper analysis, some limitations also exist in our meta-analysis. First, in most overall and subgroup analyses, moderate heterogeneity was detected and might have potential impact in the pooled results. Due to the limited information, subgroup analyses and meta regression according to other confounding factors were not performed. Second, sensitivity analysis limited to English-language studies suggested that a potential language bias was possible. However, we only included studies published in Chinese and English, more studies published in other languages should be concerned. Finally, lack of individual participants' data has restricted further adjustments by other covariables, such as ACE I/D polymorphism is unlikely to be a major determining factor in the development of SAD.In conclusion, given that all significant associations could not pass the FPRP analyses and no significant association was detected in the high quality studies, the large sample size studies and the population-based studies, we suggest that the Figure S1Cumulative meta-analysis of the relation between ACE I/D polymorphism and risk of SAD (I vs D). Each study was used as an information step. The vertical dotted line is the summary odds ratio. Bars represent 95% confidence interval (CIs)(TIF)Click here for additional data file.Figure S2Galbraith plots of association between I/D polymorphism and SAD risk. Each number represents a separate study for the indicated association and the number is the number of the respective study included into the meta-analysis (shown in shown in . (a) all(DOC)Click here for additional data file.File S1ACE I/D polymorphism and SAD risk. (2) Table S2 The univariate meta-regression results of the association of the I/D polymorphism and risk of SAD. (3) Table S3 Results of Begg’s tests and Egger’s tests for overall analyses.(1) Table S1 Scale used for quality assessment of studies of the association between (DOC)Click here for additional data file.Checklist S1PRISMA Checklist for systematic review and meta-analysis.(DOC)Click here for additional data file.Checklist S2Genetic association studies checklist.(DOCX)Click here for additional data file.


The objective was to perform a meta-analysis to summarize the available evidence from prospective nested case-control studies on the association of vitamin D receptor (VDR) polymorphism and the risk of breast cancer.We searched PubMed, ISI web of science, EMBASE, and reference lists for included articles. Study specific odds ratios (ORs) and 95% confidence intervals (CIs) were pooled by using fixed-effect or random-effects models.Eight studies were included in the meta-analysis. There were no association between Fok1 gene allele contrast f versus F , ff versus FF , recessive models ff versus FF+Ff , and dominant models ff+Ff versus FF . The estimated VDR polymorphism showed no significant association between Bsm1, Taq1, Apa1 polymorphism, and breast cancer risk. In the Caucasian ethnic subgroup, no association was found between allele contrast, recessive models, and dominant models on Fok1, Bsm1 polymorphism, and breast cancer risk.VDR polymorphism  were not associated with the risk of breast cancer in the general population as well as Caucasian population. The results were controversial. For example, Sinotte et al, Gapska et al , and McKay et al reported increased risk among ff carriers on Fok1.For consideration case-control studies are prone to selection bias. To overcome the shortcomings of the retrospective studies, we perform meta-analysis on prospective studies.We systematically searched 3 databases: PubMed, ISI web of science and EMBASE for studies published in any languages . The searched terms used are as follows: VDR; vitamin D3 receptor; vitamin D3 receptors; 1,25-dihydroxyvitamin D3 receptor; 1,25-dihydroxyvitamin D3 receptors; calcitriol receptor; calcitriol receptors; cholecalciferol receptor; cholecalciferol receptors; 1,25-dihydroxycholecalciferol receptor; 1,25-dihydroxycholecalciferol receptor; VDR combined with breast cancer; breast carcinoma; breast neoplasm; breast neoplasms; breast tumor; breast tumors; mammary cancer; mammary carcinoma; mammary carcinomas; mammary neoplasm; mammary neoplasms; mammary tumor; mammary tumors; FokI; BsmI; ApaI; TaqI; Cdx2, and polyA. The search was restricted to studies of human participants. We also have reviewed the reference lists of enrolled articles to identify additional articles. Ethical approval was not necessary. Because this was a meta-analysis it involved no direct handing of personal data or recruitment of subjects.P >0.05). Reviews, retrospective case control studies or studies with insufficient data were excluded. When there were multiple published reports from the same study population, the most recent or the most informative report was selected for analysis.For inclusion, the studies had to have met the following criteria: (1) breast cancer cases were medically confirmed pathologically; (2) the study was designed as prospective nested case-control or cohort study; (3) providing the data of VDR gene polymorphism and incidence of breast cancer; (4) detailed data of odds ratios (OR) with 95%CI; (5) all of the cases were adult; and (6) all SNPs were in Hardy–Weinberg equilibrium (HWE) , and adjustments for potential confounding. When the studied population was >95% Caucasian, we included this study in the Caucasian group.18 With regard to assessment on amount of evidence, category A required a sample size >1000, category B corresponded to a sample size of 100 to 1000, and C corresponded to a sample size <100. Sample size referred to the total number of cases and controls when the least frequent genotype was used. To assess replication, when I2 <25% category A was given, B for 25% ≤ I2 ≤50% and C for I2 >50%. To assess protection from bias, all of the following criteria were required: (1) clear phenotype definition; (2) reliable genotyping test method; and (3) no change of result when the small sample study was excluded.We used Venice interim criteria to access the cumulative evidence of the genetic association between the VDR polymorphism  and the risk of breast cancer. The Venice interim criteria included in amount of evidence, replication of results, and protection from bias.For each study, the HWE of SNPs in the control group by using the chi-square test was reported by original article. The multivariate adjusted ORs and 95% CIs presented in the literature were used.I2 statistic.19 When I2 was from 0% to 40% along with P >0.10, the heterogeneity might not be important. If the meta-analysis has no heterogeneity, fixed-effects model with the Mantel–Haeszel method20 would be used to combine the individual studies, otherwise, the random-effects method21 was used for pooling.Statistical heterogeneity among studies was tested with the Cochrane Q statistic, and statistical inconsistency was quantified with the 22 and Begg–Mazumdar test23 were used to assess for publication bias. P <0.05 was considered statistically significant publication bias.The Egger's regression testP-values were two sided. All statistical analyses were performed using STATA .All reported 8 Eight independent nested cast-control studies were eligible for our meta-analysis.24 The flow diagram of our systematic literature search is shown in Figure A total of 248 abstracts were retrieved though PubMed, ISI Web of Science, and EMBASE. After removing duplication, 205 abstracts remained. Of these 205 abstracts, we identified 51 potentially relevant studies that described the association between the VDR gene polymorphism and breast cancer risk after screening the titles and abstracts. For further assessment, 47 articles were excluded after full-test review. Nine articles were excluded because they did not investigate the association between VDR gene polymorphism and breast cancer risk, 2 were duplicate reports on the same study population, and 36 were not nested case-control studies. The study of Mckay et al contained 5 different populations, thus, it was extracted into 5 individual studies.24 study because the study population was the same as the Mckay et al's8 study. This meta-analysis included 8 prospective studies. Among these articles, all studies were conducted in the United States. Seven studies on Fok1, 6 son Bsm1, 2 on Taq1, and 2 on Apa1 were included in the meta-analysis. Ethnic subgroups were also reported on: 5 studies reported on Caucasians, 1 study on Hispanics, 1 on African Americans, 1 on Asians, and 1 on Hawaiians.The characteristics of the eligible studies in this meta-analysis are summarized in Table The results of the associations between the Fok1, Apa1, Bsm1, and Taq1 polymorphisms and the breast cancer risk are shown in Table Meta-analysis of the 7 studies suggested that there was no association between allele contrast f versus F , homozygote model ff versus FF , recessive models ff versus FF+Ff , and dominant models ff+Ff versus FF   studied the Taq1 and Apa1 polymorphisms. In the meta-analysis, the summary estimated for VDR polymorphism showed no significant association between Taq1 and Apa1 polymorphisms and breast cancer risk.P >0.05 was observed among most of the genetic models of 4 polymorphisms except for Fok1, f versus F  have been identified that may influence cancer risk.32 Over the last two decades, a number of large population-based studies were carried out to investigate the association of variants in the VDR gene polymorphism and the risk of breast cancer. However, the results of these studies are controversial. The results of our meta-analysis from 8 prospective nested case-control studies indicated that there is no association between the SNPs in VDR  and risk of breast cancer both in mixed races and Caucasian population.The pathogenesis of breast cancer remains unknown. It involves environmental factors, molecular signaling pathways, and host genetic factors. There is consistent epidemiologic evidence that increased vitamin D intake is associated with reduced risk of colorectal34 However, Zhang and Song, Wang et al, and Tang et al reported that ff genotype of Fok1 is a risk factor of breast cancer .37 The reasons are as follows: first, Huang et al is an updated and more carefully selected study than Wang et al and Tang et al.37 Second, the meta-analysis of Zhang and Song divided the article Mckay et al into 6 independent studies35 and our analysis of Mckay et al only included 5 different populations because one was the overlapping data.8 What's more, the Zhang and Song included both Mckay et al and Chen et al, which were from the same cohort and contained overlapping data.8,38 Third, our meta-analysis combined the prospective studies to overcome the shortcoming of retrospective studies on study population selection bias. Our study results in no association between Bsm1, Taq1, and Apa1 polymorphisms and breast cancer risk in mixed races. Previous meta-analysis's pooled ORs were similar to ours.39 Our study only pooled the prospective studies that are more reliable.In our meta-analysis, Fok1 polymorphism showed no association with breast cancer risk. The result is consistent with the previously meta-analysis, such as Huang et al and Xu et al.40 In subgroup analyses, we conducted meta-analysis in Caucasian population. The results remained the same. There is no observed association between Fok1 and Bsm1 polymorphisms and breast cancer risk.It is well established that VDR genotypes vary widely by ethnicity.Our study had some strong points. First, all of the previous meta-analyses on the association collected both retrospective and prospective studies. In order to reduce the likelihood of selection bias our meta-analysis enrolled prospective studies only. Second, all controls’ SNPs genotype distributions were in HWE. Third, all utilized studies were strictly consistent with inclusion criteria. Fourth, no publication bias was observed indicating that the results might be unbiased.I2 was larger than 40%, we used the random-effects method21 for pooling in order to enhance the credibility of the results. Fourth, because of limited published data, our results need to be considered with caution.As in any study, some limitations of this study should be considered. First, variant adjusted factors of ORs in each study such as age, age at menarche, menopausal status, body mass index, hormone replacement treatment usage, family history, race, smoking, etc. were different from the original studies. These gene–environment interactions could bring bias and heterogeneity in our study. Therefore, a more precise analysis should be conducted if individual data are available, which could permit the same adjusted factors. Second, significant heterogeneity was observed in overall comparisons and subgroup analyses. Selection bias, although no publication bias was observed, is a possible major source of heterogeneity. Different background and variant adjusted factors of controls might be the main reason. Third, some “weak” evidences were concluded from Venice interim criteria. All those “weak” evidence were because of category C on replication. Therefore, in our meta-analysis, when In conclusion, our study provides the evidence that VDR polymorphism  was not associated with the risk of breast cancer in general population as well as the Caucasian population. Further studies are necessary to clarify these results.