XRCC1), X-ray repair cross complementing group 3 (XRCC3), and xeroderma pigmentosum complementation group D (XPD) repair genes may lead to genetic instability and leukemogenesis. The purpose of the study was to evaluate the association between XRCC1 Arg399Gln, Arg280His and Arg194Trp, XRCC3 Thr241Met, and XPD Lys751Gln polymorphisms and the risk of developing CML in Romanian patients. A total of 156 patients diagnosed with CML and 180 healthy controls were included in this study. We found no association between CML and XRCC1 or XRCC3 variant genotypes in any of the investigated cases. A significant difference was observed in the variant genotype frequencies of the XPD Lys751Gln polymorphism between the patients with CML and control group . This was also observed when analyzing the variant 751Gln allele . Our results suggest that the XPD Lys751Gln variant genotype increases the risk of CML.The genetic polymorphisms of X-ray repair cross complementing group 1 ( ABL) gene on chromosome 9 with the breakpoint cluster region (BCR) gene on chromosome 22 [Chronic myeloid leukemia (CML) is a myeloproliferative neoplasm characterised by the Philadelphia chromosome (Ph), a reciprocal chromosomal translocation t q34;q11) leading to the fusion of the Abelson murine leukemia , nucleotide excision repair (NER), and double strand break repair (DSB repair) are the most important DNA repair pathways . MutatioCommon genetic polymorphisms in DNA repair genes might affect protein function and thus the capacity of repair DNA damage, which in turn could lead to genetic instability and leukemogenesis. Polymorphisms in DNA repair genes are thought to be a risk factor for cancer as a result of increased rate of mutations.XRCC1), X-ray repair cross complementing group 3 (XRCC3), and xeroderma pigmentosum complementation group D (XPD) have been studied extensively.Among them, polymorphisms of X-ray repair cross complementing group 1 . BER is one of the four major DNA repair pathways .The nucleotide excision repair (NER) pathway is responsible for repair of lesions such as bulky adducts and thymidine dimers . Double  XRCC1, XRCC3, and XPD genes have been identified. Among them, XRCC1 Arg399Gln, Arg280His, and Arg194Trp, XRCC3 Thr241Met, and XPD Lys751Gln polymorphisms are the most studied in cancers, including leukemia.Several single-nucleotide polymorphisms (SNPs) inXRCC1) is one of the most important genes involved in DNA repair, specifically in the base excision repair pathway and in single-strand break repair activity [ XRCC1 gene encodes a protein that is associated with DNA polymerase beta, DNA ligase III, and poly ADP-ribose polymerase (PARP) and functions in a complex to facilitate the repair of the damaged bases produced by endogenous or exogenous factors. XRCC1 Arg194Trp, Arg280His, and Arg399Gln single-nucleotide polymorphisms have been shown to have functional significance and could alter XRCC1 function, decrease the kinetics of repair mechanism, and influence susceptibility to cancer [X-ray cross complementing gene 1  [Because theleukemia –14. Alsoia (AML) , 15, 16,ia (AML) –19, chroia (AML) , 21, andia (AML) –26. XRCC1 gene polymorphisms in CML was investigated in only two studies [ XRCC1 Arg399Gln polymorphism and CML [ XRCC1 codons 194 and 399 with CML. However, this was not the case for codon 280 [The role of studies , 27. Oneodon 280 . XRCC3 gene product plays an important role in homologous recombination repair of DNA double strand breaks. XRCC3 Thr241Met gene polymorphism could be associated with impaired function of repair, because this polymorphism consisting of Met to Thr substitution might influence the enzyme's function by removing a phosphorylation site [Theion site . XRCC3 gene has been studied in association with leukemia. Yan et al. found a significant association between XRCC3 Thr241Met polymorphism and leukemia, in Asian patients [ XRCC3 Thr241Met polymorphism might be associated with AML risk. Seedhouse et al. reported no effect for the variant XRCC3 241Met gene alone in either de novo AML or therapy-related AML (t-AML) but demonstrated an increased risk of AML when both variants RAD51 135C and XRCC3 241Met alleles were present [Thepatients . Qin et patients  reported present . XPD gene (xeroderma pigmentosum group D) is involved in the nucleotide excision repair (NER) pathway. The XPD gene encodes a DNA helicase, essential for transcription initiation, nucleotide excision repair, cell cycle control, and apoptosis. Mutations in XPD gene reduce helicase activity and cause defects in NER pathway [ XPD gene, such as Arg156Arg, Asp312Asn, and Lys751Gln, have been studied in relation to lung cancer [ XPD Lys751Gln polymorphism has been investigated in different hematological malignancies, such as acute myeloid and lymphoblastic leukemia, but with contradictory results [The pathway , 33. Sing cancer  and colog cancer , 34. In  results –41. XPD Lys751Gln polymorphism are associated with low DNA repair capacity for benzo(a)pyrene adducts and UV DNA damage [ XRCC3 Thr241Met and XPD Lys751Gln gene polymorphisms in CML.There is evidence that variant homozygous genotypes ofA damage . To our  XRCC1 Arg399Gln, Arg280His, and Arg194Trp, XRCC3 Thr241Met, and XPD Lys751Gln polymorphisms because they were the most studied and have been shown to be responsible for a suboptimal DNA repair capacity. Thus, they might influence susceptibility to cancer.We focused in particular on the XRCC1 Arg399Gln, Arg280His, and Arg194Trp polymorphisms and the risk of developing CML in Romanian patients. In addition we assessed whether there was an association between XRCC3 Thr241Met and XPD Lys751Gln polymorphisms and CML, as such data are lacking from the literature.The aim of our study was to evaluate the association betweenThe study was performed with approval from the Ethics Committee of the University of Medicine and Pharmacy Tirgu Mures, Romania. The study was carried out according to the guidelines of the Declaration of Helsinki and informed consent was obtained from each participant.A total of 156 previously untreated adult patients aged 20 to 78 years diagnosed with CML  and 180 control individuals  were included in this study. The CML patients were consecutively hospitalized and diagnosed in the Hematology Clinics from Tirgu Mures and Cluj-Napoca between December 2010 and December 2013, according to current WHO standards , 44. ConThe median hemoglobin (Hb) level at diagnosis was 9.86 g/dL . The median blasts percentage in peripheral blood at presentation was 9.48%. Additional cytogenetic abnormalities (ACA) were observed in 20 CML cases (12.8%). There were 134 patients (85.9%) in chronic phase, 8 patients (5.1%) in accelerated phase, and 14 patients (9.0%) in blast phase.Patients received first-line therapy with imatinib mesylate (Gleevec), 400 mg/day in chronic phase. In the case of suboptimal response and failure to imatinib treatment the dose was increased to 600 mg/day imatinib, or they received dasatinib (Sprycel) or nilotinib (Tasigna). XRCC1 genotypes were determined by the polymerase chain reaction-restriction fragment length polymorphism (PCR-RFLP) assay. The primers, restriction enzymes, and PCR conditions for XRCC1 were the same as described by Batar et al. [ XRCC3 Thr241Met genotypes were detected using a PCR-RFLP method, as described previously [ XPD Lys751Gln polymorphism was also investigated by a PCR-RFLP assay, as described by Seedhouse et al. [Genomic DNA was obtained from peripheral blood samples using the commercially available Quick-gDNA MiniPrep kit  and Wizard Genomic DNA Purification Kit . Ther et al. , Seedhour et al. , and Wanr et al. . XRCC3 Teviously . The XPDP value less than 0.05 was considered as statistically significant. The odds ratio (OR) and 95% confidence intervals (CIs) were used to estimate the strength of the association between alleles and genotype in CML patients and controls. Moreover, the Hardy-Weinberg equilibrium was evaluated using chi-squared test.All data were analyzed by GraphPad InStat software, version 3 . Fischer's exact test (two-sided) was used to compare the distribution of qualitative variables between cases and controls. A The observed genotype frequencies in controls were consistent with the Hardy-Weinberg equilibrium. XRCC1, XPD, and XRCC3 gene polymorphisms are summarized in The genotype distribution and the allele frequency of the five polymorphisms analyzed are shown in  XRCC1 and XRCC3 variants. In the case of the three XRCC1 polymorphisms analyzed, the distribution of the variant heterozygous and homozygous genotypes was similar in patients and controls.We did not observe an association between CML and XRCC3 Thr241Met polymorphism, the heterozygous and variant homozygous genotypes shared similar frequencies in CML patients and controls. Also, the variant allele frequencies were similar in patients and controls in the case of all XRCC1 and XRCC3 polymorphisms analyzed.Similarly, in the case of the XPD Lys751Gln polymorphism and CML was noted. A statistically significant difference was observed in the variant genotype frequencies of the XPD Lys751Gln polymorphism between the patients with CML and control group . This was also observed when analyzing the variant 751Gln allele .In this study, an association between XRCC1, XRCC3, and XPD variants in patients stratified by gender. The XRCC1 Arg194Trp, XRCC1 Arg280His, XRCC1 Arg399Gln, XRCC3 Thr241Met, and XPD Lys751Gln polymorphisms studied had no influence on the risk of CML with respect to gender.We analysed the distribution of XRCC1, XRCC3, and XPD polymorphisms regarding blasts and white blood cells count . When the Sokal and Hasford risk groups were considered, no association was seen between variant genotypes for the XRCC1, XRCC3, and XPD polymorphisms and the risk groups mentioned above.We evaluated the impact of these polymorphisms in more detail taking into account different prognostic factors. In the current study, no association was observed in the distribution of any of the XRCC1 Arg194Trp, XRCC1 Arg280His, XRCC1 Arg399Gln, XRCC3 Thr241Met, and XPD Lys751Gln polymorphisms regarding ACA .Finally, we performed a comparison of the ACA with respect to the studied polymorphisms. No association was seen in the distribution of theP value > 0.05 for all these comparisons).Patients with specific genotypes were not more likely to receive a particular tyrosine kinase inhibitor , for none of the polymorphisms analyzed  have a higher risk of developing AML. Similar results were reported by Joseph et al. [El-Din et al.  observedh et al.  in patie XRCC1 variant 280His allele is associated with a reduced capacity of single-strand breaks (SSB) and BER systems, which consequently increases the risk of carcinogenesis. However, our study did not reveal a statistical significant difference between CML patients and controls, regarding the distribution of the XRCC1 Arg280His polymorphism.Takanami et al.  reported XRCC1 Arg399Gln, Arg194Trp, and Arg280His polymorphisms and leukemia risk. The findings of the meta-analysis demonstrate that XRCC1 Arg399Gln, Arg194Trp, and Arg280His polymorphisms are not associated with overall leukemia risk, but they could be associated with the risk for some specific leukemia entities [Our results are similar to that observed by Zhang et al. in a recent meta-analysis of 19 case-control studies which evaluated the association betweenentities . XRCC1 polymorphisms may be due to the ethnic origin, sample size of the studied populations, and different study designs. Also, variation in carcinogenic exposure, alcohol consumption, and cigarette smoking may contribute to differing results.The contradictory results from different studies performed on XRCC1 194Trp, 399Gln, and 280His alleles in our CML patients were 0.13, 0.32, and 0.26, whereas in controls they were 0.16, 0.30, and 0.21, respectively. Deligezer et al. [ XRCC1 Arg399Gln polymorphism on a cohort which included 182 cases of CML and 226 controls from Turkey. The frequency of the variant Gln allele was 0.35 in controls and 0.34 in CML cases. In a recent study Annamaneni et al. [ XRCC1 repair gene  with CML in 350 patients from Hyderabad, India (South Asia). In the study mentioned above, the frequency of XRCC1 Gln, His, and Trp alleles was 0.50, 0.006, and 0.85 in CML patients, whereas it was 0.49, 0.018, and 0.81, respectively, in controls [ XRCC1 399Gln allele and its distribution in the control group were similar to those found in the population from Turkey [ XRCC1 194Trp allele was higher, while the frequency of the 280His allele was similar in controls from India [The frequencies ofr et al.  analyzedi et al.  exploredcontrols . Thus, tm Turkey  and lessm Turkey , suggestom India , compare XRCC1 polymorphisms do not only increase the susceptibility to CML but also may predispose to developing ACA in CML. When comparing patients with to those without ACA, genotype frequencies of the investigated polymorphisms were not found to be significantly different.We supposed that XRCC1 polymorphisms regarding age, gender, and Sokal and Hasford risk groups when comparing wild-type genotypes with variant genotypes. However, we observed an increased frequency of the XRCC1 Arg194Trp polymorphism among CML patients in accelerated and blast phase. This observation attained a borderline statistical significance (P = 0.05).In the current study, no association was seen in the distribution of the XRCC3 Thr241Met polymorphism in our patients with CML and controls. Our results suggest that the XRCC3 Thr241Met variant genotype is not a risk factor for the development of CML. No association was observed between the prognostic factors  and the XRCC3 Thr241Met variant genotypes in patients with CML.We also studied the genotype distribution of the XRCC3 Thr241Met polymorphism and leukemia risk in overall populations, but significant association between XRCC3 Thr241Met polymorphism and leukemia risk was found in Asians [Similar results were reported by Yan et al. in a meta-analysis which included seven studies with 1070 cases and 1850 controls . Yan et n Asians . XRCC3 Thr241 Met polymorphism and the risk of ovarian cancer [No significant association was found between then cancer . XRCC3 genes polymorphisms might play an important role in the development of AML.These findings are not in agreement with the studies conducted by Voso et al.  and Hamd XRCC3 241Met allele was 0.36 in CML patients and 0.30 in controls. According to Seedhouse et al. [ de novo AML and 186 controls, the variant allele frequencies were 0.29 in controls and 0.34 in AML patients. In the study of Voso et al. [ XRCC3 241Met allele was 0.45 in AML patients from Italy. Thus, the frequencies for the XRCC3 241Met allele were similar to those found in other Caucasian populations [In our study, the frequency ofe et al. , which io et al.  the frequlations , 51. XPD Lys751Gln polymorphism and CML risk. Our results suggest a positive association between the XPD Lys751Gln variant homozygous (Gln/Gln) and combined heterozygous + homozygous variant genotypes (Lys/Gln + Gln/Gln) and the risk of CML. We also observed an association between variant XPD 751Gln allele and the risk of CML. These results suggest that the XPD Lys751Gln polymorphism may contribute to leukemogenesis in CML. Our findings are in agreement with the study conducted by Özcan et al. [ XPD 751Gln allele is associated with a reduced DNA repair capacity and increased leukemogenic risk and that XPD Lys751Gln polymorphism may affect the outcome of childhood AML therapy [ XPD Lys751Gln variant genotypes were shown to be risk factors for AML [ XPD Lys751Gln polymorphism between AML patients and controls [We also evaluated the potential role ofn et al. . They su therapy . Similar for AML  and acut for AML . These fcontrols . XPD 751Gln allele was 0.42 in our CML patients and 0.32 in controls. According to Wang et al., the frequency of the XPD 751Gln allele was 0.39 in Europeans, 0.36 in Americans, 0.12 in Asians, and 0.24 in Afro-Americans [ XPD codon 751 was associated with increased cancer risk compared with the Lys/Lys genotype only in the European population [ XPD 751Gln allele frequency was 0.38 among the controls, while it was 0.30 in AML patients from Egypt [The frequency ofmericans . In a mepulation . The varom Egypt . XPD Lys751Gln polymorphism increases the risk of CML. According to our findings, the XRCC1 Arg194Trp, Arg280His, Arg399Gln, and XRCC3 Thr241Met polymorphisms are not a risk factor for CML.In conclusion, our study suggests that the XRCC1, XRCC3, and XPD polymorphisms and CML.In the future, similar studies performed on larger cohorts of patients should clarify the relationship between


Association between dietary intake of vegetables and fruits and risk of hip fracture has been reported for many years. However, the findings remain inconclusive. We conducted a meta-analysis to evaluate the relationship between intake of vegetables and fruits, and risk of hip fracture. Literature search for relevant studies was performed on PubMed and Embase databases. Five observational studies were included in the meta-analysis. Summary hazard ratio (HR) with corresponding 95% confidence interval (CI) was calculated from pooled data using the random-effects model irrespective of heterogeneity. Sensitivity and subgroup analysis were performed to explore possible reasons for heterogeneity. The summary HR for hip fracture in relation to high intake vs. low intake of only vegetables, only fruits, and combined intake of fruits and vegetables, was 0.75 , 0.87 , and 0.79 , respectively. Subgroup analyses based on study design, geographical location, number of cases, and gender showed similar results. Increased intake of vegetables, but not fruits, was found to be associated with a lower risk of hip fracture. Large prospective clinical trials with robust methodology are required to confirm our findings. Osteoporosis and osteoporotic fractures account for a considerable proportion of disability, morbidity and mortality worldwide, especially in the elderly population, which imposes a significant social and economic burden1The etiopathogenesis of hip fracture is complex812131619201722The descriptive data for all included studies are summarized in 1214151214131213141512As shown in the P = 0.028). However, significant heterogeneity was observed between the studies     . SensitiP = 0.119) using the random-effects model, with a high heterogeneity    using a random-effects model, with a high heterogeneity   . Calcium is a key element required to maintain bone health. The association of potassium and magnesium with bone health is well-documented34To the best of our knowledge this meta-analysis is the first to investigate the association between intake of fruits and vegetables and the risk of hip fracture. The large sample size and a longer time span serve to enhance the statistical power to detect possible associations. The power of HR value of intake of only fruits, and only vegetables with risk of hip fracture was 83.48% and 98.43%. Moreover, the use of multivariable-adjusted risk estimates minimized the confounding factors.However, the potential limitations of this meta-analysis ought to be taken into account while interpreting the results. First, we cannot exclude the possibility that the observed inverse association between intake of vegetables and hip fracture risk could be owe to unmeasured or residual confounding. Further, though the meta-analysis involved a large number of participants, none of the included studies were interventional studies. Third, although we chose the highest multivariable adjusted effect estimates in our meta-analysis, we cannot rule out the effect of uncontrolled confounding by imprecise quantification of the dietary intake of fruits and vegetables. Fourth, a significant heterogeneity was detected among the studies. We further performed sensitivity and subgroup analyses to determine the sources of heterogeneity. The observed heterogeneity could partly be explained by study design (case-control studies and cohort studies) and the articleIn summary, our results showed that intake of vegetables, but not fruits, may be associated with a lower risk of hip fracture. However, prospective studies need to be conducted to confirm the findings of this meta-analysis.A comprehensive literature search for relevant studies published up to May 2015 was performed in PubMed and EMBASE databases. The following search terms were used without imposing any limitation: “fruit(s)” or “vegetable(s)” or “brassicaceae” or “cruciferae” or “citrous” or “citrus”, combined with “hip fracture(s)”. The reference lists in the retrieved publications were manually screened to increase the yield of relevant studies.Studies were considered eligible for inclusion if they met the following criteria: 1) study design: either cohort, case control or cross-sectional; 2) the exposure variable was fruit and/or vegetable consumption; 3) study outcomes included hip fracture risk; 4) Hazard ratio (HR)/Relative risk (RR) and Odds ratio (OR)  reported . The exclusion criteria were editorial letters, historical reviews and descriptive studies, such as case reports and case series, or laboratory studies. When multiple publications covered the same study population, only the study with the larger sample size was included. Two authors (JKS and HXY) independently performed the literature search and reviewed all studies for their eligibility for inclusion in the meta-analysis. Any disagreements were resolved by consensus or after consultation with the third author (GLH).Data pertaining to the following variables were independently extracted for each study by two authors (JKS and HXY) using a standardized data extraction form: first author’s surname, publication year, study design, country, number of subjects (cases/participants), gender, exposure type, methodology for dietary assessment, and covariates included in the multivariable model. Disagreements were resolved by discussion and a consensus reached by involving the third author (GLH), if necessary.The methodological quality of the included studies was evaluated using the Newcastle-Ottawa scale (NOS)et al.Hazard ratio with 95% CI was the common measure across all eligible studies. Since hip fracture is a relatively rare event, the differences among risk estimates  were ignored, and the OR and RR were directly converted into HR. A random-effects model of DerSimonian and Laird was used to calculate summary HR, comparing the highest across all included studies regardless of heterogeneitySensitivity analysis was performed to evaluate robustness and stability of the results by sequentially omitting one study on each turn. Moreover, subgroup analysis was performed to assess heterogeneity and assess the influence of different inclusion criteria on the overall estimate.How to cite this article: Luo, S. et al. Increased intake of vegetables, but not fruits, may be associated with reduced risk of hip fracture: A meta-analysis. Sci. Rep. 6, 19783; doi: 10.1038/srep19783 (2016).


Several case–control studies have been conducted to clarify the association between the tumor necrosis factor alpha (TNF-α) -G308A polymorphism and risk of osteoarthritis (OA); however, the results are inconsistent. This meta-analysis was performed to clarify this issue using all the available evidence.Eligible articles were retrieved by searching PubMed, Web of science and Google scholar. The strength of the association between the TNF-α -G308A polymorphism and risk of OA was assessed by odds ratios (ORs) with the corresponding 95% confidence interval (CI) for each study.vs. G allele analysis . However, there was no statistical difference in the dominant genetic model analysis . Furthermore, we found that OA patients had a higher frequency of the AA genotype  and GA genotype  compared with the control population.Seven studies were included in the meta-analysis, which included 983 OA cases and 1355 controls. The pooled analysis based on all included studies showed a significantly increased OA risk in the recessive genetic model analysis  and in the A allele Our results suggested that the TNF-α -G308A polymorphism were associated with an increased risk of OA.The online version of this article (doi:10.1186/1471-2474-15-373) contains supplementary material, which is available to authorized users. As one of the most common age-related disease leading to restrictions in daily activities of the elderly, osteoarthritis (OA) is set to become the 4th-highest impact condition in women and the 8th-highest in men. The prevalence of OA is steadily increasing across all age groups . Given tTNF-α is a multifunctional pro-inflammatory cytokine involved in various physiological and pathological processes, including inflammation, immunoregulation, proliferation and apoptosis . TNF-α iGenetic variations within the TNF-α promoter could alter the transcription and translation of TNF-α . PolymorEligible articles were retrieved by searching the PubMed bibliographical database, Web of science and Google scholar  using the following combination of keywords:  OR (tumour necrosis factor-α) OR (TNF-α)); OA OR osteoarthritis; polymorphism OR polymorphisms OR variants OR variant. In addition, we checked the references in reviews and in the retrieved articles to avoid missing any relevant studies. There was no restriction on language in the search. The article selection was carried out by both YW and SK independently to minimize bias and human error.To be included in this meta-analysis, an article had to provide the following information: 1) the number of OA cases and controls; and 2) the number of individuals with GG, GA and AA in both OA cases and controls. Those not designed as case–control studies, systemic reviews, and those that provided no usable data were excluded.Two independent reviewers used a predesigned data extraction table to extract the data. The following information was extracted from each included article: journal name, first author, publication year, ethnicity and country, OA phenotype, source of controls, the number of genotypes in OA cases and controls, and the results of the study. Disagreement was resolved by discussion.vs. GG) ; 2) the recessive genetic model (AA vs. GA + GG) and 3) the allele A vs. allele G analysis. The genotype frequencies of GG, GA and, AA were also assessed using the same method. A chi-squared (χ2) test was used to assess heterogeneity across studies. A fixed effect model was used when no heterogeneity existed among the studies. Otherwise, a random effect model was used. Meta-regression analysis was performed to find any sources of heterogeneity. Subgroup analysis for ethnicity (Asian and Caucasian), design type  and PCC (population based case–control study)) and sample size  and larger ) was conducted. Potential publication bias was assessed using a funnel plot, and calculated using Begg’s and Egger’s tests (P < 0.05 was considered as significant publication bias). Influence analysis was performed by omitting each study to find potential outliers. In the control populations, the Hardy-Weinberg equilibrium (HWE) was tested using a chi- squared (χ2) test. Two authors performed the statistical analysis independently and obtained the same results. STATA software  was used to perform the statistical analyses. P values less than 0.05 were considered statistically significant.The strength of the association between the TNF-α -G308A polymorphism and risk of OA was assessed by odds ratios (ORs) with the corresponding 95% confidence interval (CI) for each study. The ORs and their 95% CIs were assessed for the following genetic models: 1) the dominant genetic model . Fourteen articles were selected for detailed assessment and seven of them were excluded . Finally, seven studies were included in this meta-analysis, which included 983 OA cases and 1355 controls. The detailed selection procedure is shown in Figure vs. G allele analysis   . A random effects model was adopted in these two analyses. Furthermore, a meta-regression analysis was conducted to find potential sources of heterogeneities. Unfortunately, the common variables, such as publication year, ethnicity, study design type and sample size were not significant sources of heterogeneities. However, we still performed subgroup analysis based on ethnicity, study design type  and total sample size, because such subgroup analysis was valuable. The detailed results are shown in Table Influence analysis was performed to assess the sensitivity of each individual study on the pooled ORs, by sequential omission of each individual study. The results suggested that Han’s study  significFunnel plots were used to estimate qualitatively the potential publication bias. Taking the the recessive model analysis and A allele vs. G allele analysis as examples, the shapes of the funnel plots did not indicate any obvious asymmetry Figure A,B. FurtTo the best of our knowledge, this is the first meta-analysis to evaluate the association between the TNF-α -308 polymorphism and risk of OA. In this meta-analysis, we discovered an increased OA risk in the recessive genetic model analysis and in the A allele vs. G allele analysis. Furthermore, we found a higher frequency of the AA and GA genotype in OA patients.vs. GG analysis and the recessive model analysis, and higher frequencies of the AA and GA genotypes in OA patients. However, by subgroup analysis, we found that the relationship of TNF-α-308 polymorphism and OA risk only existed among Asian population but not among Caucasion population. In recent GWAS studies, 11 OA susceptibility loci with genome-wide significance levels was indentified. Among them, a region containing HLA class II/III genes, which close to the region habouring TNF gene, showed association in Asians but not in Europeans [TNF-α is a potent pro-inflammatory cytokine in inflammatory diseases and the immune response. Higher levels of TNF-α were found in OA patients compared with the control population . TNF-α curopeans –30, otheuropeans . These fInfluence analysis suggested that Han’s study  might haRecently, we have noticed that some studies have been conducted to explore the efficacy of anti-TNF therapy on OA. Güler-Yüksel’s study have found that treatment with TNF-α inhibitor might reduce hand OA in patients with rheumatoid arthritis, which sAlthough the primary results of our meta-analysis are suggestive, some limitations may still exist. Firstly, there was heterogeneity between studies of the -308 polymorphism and the heterogeneity may have distorted the meta-analysis. Unfortunately, meta-regression analysis failed to find the potential heterogeneity. In addition, no quality tool has been used to assess the risk of bias in the included studies, study quality could be one factor missing from the heterogeneity analysis. Secondly, the number of OA cases and studies included in the Caucasian-specific meta-analysis was low. This study may not have enough power to explore the association between the TNF-α -308 polymorphism and risk of OA in the Caucasian population. Thirdly, the results of the meta-analysis were based on unadjusted estimates owing to the lack of adjusted estimates. Currently, some risk factors have been identified, such as aging, obesity, previous injury, smoking habit and hormone therapy. A more precise analysis could be performed if these data were available. Finally, gene-gene interactions were not fully addressed in the meta-analysis because of the lack of relevant data. In fact, the TNF-α gene is located close to the human leucocyte antigen (HLA) class II/III region, in which some SNPs, such as rs7775228 and rs10947262, have been identified as associated with risk of knee OA . TherefoIn conclusion, this comprehensive meta-analysis has evaluated all published data currently available on the TNF-α -308 polymorphism and risk of OA. Our meta-analysis suggested that the AA and GA genotypes might increase the risk of OA compared with the GG genotype, which may be explained by the higher expression of TNF-α in the -308A allele carriers than in the -308G carriers.Additional file 1: Figure S1: Meta-analysis of TNF-α -308 polymorphism and OA risk after omitting Han’s study: (A) dominant genetic model analysis; (B) recessive genetic model analysis; (C) A allele vs. G allele analysis. (TIFF 12 MB)Additional file 2: Figure S2: Meta-analysis of TNF-α -308 genotypes and OA risk after omitting Han’s study: (A) AA genotype frequency; (B) GA genotype frequency; (C) GG genotype frequency. (TIFF 12 MB)Authors’ original file for figure 1Authors’ original file for figure 2Authors’ original file for figure 3Authors’ original file for figure 4Authors’ original file for figure 5Below are the links to the authors’ original submitted files for images.


The relationships between dietary fruit, vegetable, fat, and red and processed meat intakes and Barrett’s esophagus (BE) risk remain inconclusive. We conducted a systematic review and meta-analysis to summarize the available evidence on these issues. PubMed, EMBASE and the Cochrane Library were searched for studies published from inception through October 2015. A total of eight studies were included in this analysis. Fruit intake was not associated with BE risk , but vegetable intake was strongly associated with BE risk . Saturated fat, red meat and processed meat intakes were not associated with BE risk with OR = 1.25 (95% CI = 0.82–1.91), OR = 0.85 (95% CI = 0.61–1.17) and OR = 1.03 (95% CI = 0.73–1.46), respectively. Dietary vegetable not fruits intake may be associated with decreased BE risk. Fat and red and processed meat intakes may not contribute to an increased BE risk. Well-designed, large prospective studies with better established dose-response relationships are needed to further validate these issues. Barrett’s esophagus (BE) is considered to be the strongest risk factor2156H. pylori infection, smoking status and alcohol intake, the causes and pathogenesis of the high prevalence of BE are still unclear912131415BE is defined as the presence of columnar-type mucosa of the esophagus on endoscopy and pathology, and BE is also recognized as a specialized intestinal metaplasia by histology of the esophagus in most studies15910There are some studies17Studies unrelated to our topics were excluded.Case-control studies and cohort studies were included.The patients with BE were all diagnosed via endoscopy and biopsy. The histological feature, which was not consistent with the diagnostic gold standard, was excluded.Data that were incomplete or could not be combined were excluded.Narrative reviews, systematic reviews and meta-analyses were excluded.Cases in which only comments, case reports, editorials, letters and the abstract could be obtained were excluded.The most recent studies, the most samples, and the quality studies that included reports with the same patients were selected.Red and processed meats in this study included beef, corned beef, beefburgers, veal, bacon, bacon rashers, luncheon meat, lamb, mutton, ham, sausage, salami, hot dogs, souse meat, smoked meat, salted meat and barbecued meat.Poultry, chicken, fish and other white meats were excluded.“fruit”, “fruits”, “vegetable”, “vegetables”, “fat”, “meat”, “red meat”, “processed meat”, “beef”, “corned beef”, “beefburgers”, “veal”, “bacon”, “bacon rashers”, “pork”, “lamb”, “mutton”, “luncheon meat”, “ham”, “sausage”, “salami”, “hot dogs”, “sauce meat”, “smoked meat” “salted meat”, “barbecued meat” in combination with “Barrett’s esophagus/oesophagus”, “gastrointestinal tract”, “metaplasia”, “precancerous/neoplasia/canceration/cancerization”, “carcinogenesis/tumorigenesis”, “esophageal cancer/carcinoma/adenomas/adenocarcinoma”. Reference lists in the included studies were also searched manually to identify additional literature. The two sets of keywords were combined individually, and the eligibility criterion was independently judged by two authors (Zhanwei Zhao and Zhongshu Pu). In case of disagreement between the two authors, the third author (Qingchuan Zhao) made a consensus decision. The language of all studies was limited to English only. The studies were also limited to only those in humans.We searched PubMed, EMBASE and the Cochrane Library for studies of fruits, vegetables, fat, and red and processed meats and BE published from inception through October 2015. The following search terms were used: BE is defined as the presence of columnar-type mucosa in the esophagus on endoscopy and pathology. Subsequently, BE is also recognized as a specialized intestinal metaplasia by histology of the esophagus in most studies155In this study, red and processed meats included beef, corned beef, pork, lamb, mutton, beefburgers, veal, bacon, bacon rashers, luncheon meat, ham, sausage, salami, hot dogs, souse meat, smoked meat, salted meat and barbecued meat.Study quality in this meta-analysis was assessed using the Newcastle-Ottawa Scale (NOS)A data extraction sheet was set up to enter data from each study, including the first author, year of publication, country, study type, case/control, study population, method of dietary assessment, type of dietary exposure measured, meat exposure categories, NOS score and controlled variables . The conI2 statistic The data were collected and extracted using SPSS 17.0 . RevMan 5.3  and STATA, version 12.1  software were used for data synthesis and analysis. Heterogeneity was detected using the The 8 selected studies included 6 case-control studies and 2 cohort studies with 858 cases of fruit intake, 858 cases of vegetable intake, 667 cases of fat intake, 412 cases of red meat intake and 633 cases of processed meat intake . The NOSP = 0.004, I 2  = 77%). No statistical evidence of publication bias was observed from Begg’s test (P = 0.089). These results suggest that fruit intake may be not significantly associated with BE risk.The ORs have been adjusted for the potential multivariable confounders, including age, sex, ethnicity, energy intake, body mass index, waist-to-hip ratio, smoking, alcohol, education, medication use, gastro-esophageal reflux and vitamins. P = 0.05, I 2 = 61%) with no publication bias according to Begg’s test (P = 0.308). The results indicate that vegetable intake may have a strong inverse association with BE risk (with a 55% decreased risk of BE).The ORs (95% CI) of each study and the pooled OR (95% CI) for the highest versus lowest category are listed in P = 0.69, I 2 = 0%) and no publication bias according to Begg’s test (P = 0.540). There were two case-control studies for monounsaturated fat . Begg’s test showed no statistical evidence of publication bias (P = 1.000). The pooled OR of polyunsaturated fat intake was 0.67 (95% CI = 0.35–1.26) with no heterogeneity . Additionally, OR = 1.03 (95% CI = 0.48–2.23) indicated that there was no association between cholesterol intake and BE risk  with no significant heterogeneity . These results suggest that red and processed meat consumption may be not associated with BE risk.The ORs (95% CI) of each study and the pooled OR (95% CI) for the highest versus lowest category are listed in We also performed a dose-response meta-analysis to evaluate the dose-response associations between dietary fruit and vegetable1719232323The dose-response analysis showed that BE risk decreased 16%  per unit increase (serving/day) with vegetable intake. There were no significant changes of BE risk per unit increase (serving/day) with fruit intake  and per unit increase (gram/day) with saturated fat intake , monounsaturated fat intake  or polyunsaturated fat intake .et al.Studies have explored and reported the possible relationships and mechanisms supporting the idea that high fruit and vegetable intakes reduce BE risk. Dietary fruits and vegetables may be inversely associated with BE risk, which may influence the early carcinogenesis of EACrd compared with 1st)23Polyunsaturated fat and omega 3 intakes show inverse associations with BE risk that are stronger for long-segment BERecently, meats, particularly red and processed meat intakes, have been reported as risk factors for oral cavity and oropharynx cancer2830313335364042455518Although some studies have reported that dietary fruit, vegetable, fat, and red and processed meat intakes were associated with BE risk, the related evidence is sparse and inconsistent, and there had been no published meta-analysis. This study focuses on these issues for the first time and provides reliable evidence to date.Eight hundred fifty-eight cases of fruit intake, 858 cases of vegetable intake, 667 cases of fat intake, 412 cases of red meat intake and 633 cases of processed meat intake were included in this analysis. Our findings indicated that dietary vegetable rather than fruit intake was associated with significantly reduced BE risk (with 55% lower risk). Furthermore, the dose-response analysis suggested that the increases in vegetable [per unit increase (serving/day)] intake was significantly associated with a 16% decreased risk of BE.Although previous studies have reported that red and processed meat intakes were associated with a significantly higher incidence of BEOverall, these results suggest that essential moderate vegetable intake was associated with decreased BE risk, which may be important for the prevention of EAC. Dietary fruit, fat and red and processed meat intakes may be not associated with BE risk. Nonetheless, these results should be treated with caution, and more high-quality designs are needed to further validate these findings because of the limited studies included, the heterogeneity among the studies and the influence of potential confounders (listed in the following section on limitations).Because there is no previously published meta-analysis that has evaluated the overall effects of fruit, vegetable, fat, and red and processed meat intakes on BE risk, a quantitative synthesis of the eligible data from included studies was required to provide important evidence on the associations. The dose-response analysis was also conducted to further assess these associations.H. pylori and vitamins. Second, the quality of some studies was not high, despite meeting the criteria, and the sample size was not large. Third, there were five studies performed in the USA, two studies performed in the Netherlands and one study performed in the UK .


Analysis about the effect of EF on differences in participants’ energy intake revealed that increased EF was associated with higher energy intake . We conclude that increased EF may lead to lower obesity risk but higher energy intake. Clinical trials are warranted to confirm these results and to assess the clinical practice applicability.Body weight is regulated by energy intake which occurs several times a day in humans. In this meta-analysis, we evaluated whether eating frequency (EF) is associated with obesity risk and energy intake in adults without any dietary restriction. Experimental and observational studies published before July 2015 were selected through English-language literature searches in several databases. These studies reported the association between EF and obesity risk  in adults who were not in dietary restriction. R software was used to perform statistical analyses. Ten cross-sectional studies, consisting of 65,742 participants, were included in this analysis. ORs were considered as effect size for the analysis about the effect of EF on obesity risk. Results showed that the increase of EF was associated with 0.83 time lower odds of obesity ( Prevalence of overweight and obesity, which are established risk factors for non-communicable diseases such as cardiovascular diseases, diabetes, and certain cancers, has dramatically increased around the world since 1980. In 2014, 39% and 13% of adults (above 20 years old) were overweight and obese worldwide, respectively [The fundamental cause of obesity is an energy imbalance between energy intake (EI) and energy expenditure (EE). Weight gain indicates that the balance has tipped towards increased EI or reduced EE or a combination of both . In humaet al. reported that increased EF improved appetite control and subsequently reduced EI in obese males [One of the modifiable dietary behaviors is eating frequency (EF). Commonly, an eating occasion is defined as any instance in which participants reported consumption of solid meals and snacks, with a minimum gap of 15 min between two eating episodes . Howeverse males . Interesse males .et al. found that without caloric restriction on a daily basis, rats fed ad libitum gained weight faster than three-time schedule-fed rats [et al. pointed out this inverse relationship in the 1960s [et al. have shown that increased EF was associated with lower obesity risk in children and adolescents, especially in boys [et al. suggested a potential benefit of higher EF for body weight status during weight loss, although the positive findings were produced by a single study [et al. suggested that after correcting the effect of physical activity EE, the association between EF and obesity was no longer significant [Despite conclusion about the association between EF and EI is far from consistent, it is plausible that EF may influence at least one side of energy balance and thus affects body weight. Previous evidence from animal studies suggested that the effect of EF on energy balance and body composition was not obvious in rats during caloric restriction . This prfed rats . For humfed rats ,14,15. She 1960s , numerou in boys . Similarle study . Howeverle study  or positle study ,22,23 benificant . The preThis meta-analysis was carried out in accordance with the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) statement .Original research and observational studies (including those identified via review articles) published before July 2015, examining the association between EF and obesity in adults who were not during weight loss, were selected through English-language literature searches in the PubMed, Elsevier Science Direct, Nature, Science Online and Embase databases. Combinations of at least two of the following key words were used as search terms: meal frequency, feeding frequency, EF, eating patterns, eating behaviors, body composition, weight, body mass index (BMI), obesity and EI. In addition, the reference lists of the eligible retrieved articles were used to identify relevant articles that were not extracted through the searching procedure. Abstracts from conferences, reviews, and unpublished dissertations or theses were excluded from analysis. To reduce the potential for selection bias, each study was independently evaluated by 2 of the investigators (Yue-Qiao Wang and Yi-Wen Zhang), and a mutual decision was made jointly as to whether or not it met the basic inclusion criteria. The disagreements were solved by the opinion of the third author (Fei Zhang), and consensus was reached by discussion. Studies were included if they met the following criteria: (1) original research or observational studies published in an English-language refereed journal; (2) all of the selected studies were conducted in adults; (3) participants or subjects were not during weight loss and diet control; (4) participants or subjects did not have psychological and eating disorders; (5) no experience of bariatric surgery; (6) non-pregnancy; (7) the independent variable was EF; (8) the dependent variable was risk of obesity or EI; (9) results were provided in a form that could be used for the present analysis (point estimates of odd ratios (OR) with 95% confidence intervals (CI) or the differences in the mean values of the EI among EF categories and standard errors (SEs) available).The following information was extracted from each included study: first author’s last name; year of publication; country of origin; sample size; age and gender of participants, and the number of daily meals/eating episodes; effect size measurements; 95% CIs; SEs or SDs; the evaluation of EF and EI; the deﬁnition of a meal/eating episode and variables that entered into the multivariable model as potential confounding factors. Information and data was extracted by one author (Yue-Qiao Wang) and checked by another (Yun-Quan Zhang). Quality of studies was not assessed because the total number of studies was limited. Individual study quality was assessed using a score modified from the Agency for Healthcare Research and Quality (AHRQ) Cross-Sectional Study Quality Assessment . For eacversus the reference constituted the units of the meta-analysis. When more than one comparisons of different EF categories were included in the same study, an overall estimate for the study was calculated from ORs using the fixed-effect model [For studies investigating the relationship between EF and obesity risk, the effect size was presented as ORs and their corresponding 95% CIs. The lowest category was considered as reference group. If the original study used the highest category as reference, reciprocal of original ORs was calculated for analysis. Single comparison between one EF category ct model . For exaFor studies investigating the relationship between EF and EI, the effect size was shown as the differences of EI among categories and their corresponding SEs. If the study reported EI in kilojoule or megajoule, we unified the unit as Kcal. In each study, we calculated the differences of EI among categories. For example, if the EF categories included ≤3; 4–5 and ≥6, we calculated the differences of EI between each two groups as well as their corresponding SEs, and computed the overall estimate from the difference of EI using the fixed-effect model. If SEs or SDs were not reported, SEs were computed from 95% CI following a standard methodology .Q and I2, and evaluated by performing the χ2 test  [p value was less than 0.10 and I2 exceeded 50%, indicating the presence of heterogeneity, a random-effects model (the DerSimonian and Laird method) was used [If a paper reported the results of different multivariate models, the most stringently controlled estimate were extracted. Statistical heterogeneity that was attributed to studies rather than to chance was assessed by visual inspection of forest plots as well as using Cochran’s p value) . If the was used ; otherwiwas used . The prewas used ,32,33. ADetailed characteristics of selected studies were presented in et al. [et al., however, the minimum gap between two eating episodes was an hour [et al. [et al. [In order to evaluate participants’ dietary behavior, five studies collected data by self-reported questionnaires ,36,41,42et al. . To compet al. . In mostet al. ,40. Acco an hour . In four an hour ,36,39,42 [et al.  excluded [et al. ,41,42 wh [et al. ,38,39,40i.e., OR = 0.83, 95% CI 0.70–0.99, p = 0.040). A significant heterogeneity of the effect size of EF on obesity was revealed . Publication bias test by visual inspection of asymmetric plot indicated potential presence of publication bias, while the Egger test revealed no publication bias  with significant heterogeneity , which was mainly caused by the studies conducted by Karatzi et al. [et al. [p = 0.681, Analysis of the effect of EF on differences in participants’ EI was shown in i et al.  and AljuA sensitivity analysis was carried out for each model by conducting a stepwise exclusion of the results of each study. Data were not influenced excessively by omitting any single study, with the values of ORs ranging from 0.795 to 0.892 and βs of EI from 106.29 to 164.99.I2 value of 41.76%, suggesting that EF definition might be the cause of the significant heterogeneity. In addition, subgroup analyses performed on age (>40 years) and social economic status (adjusted) had lower I2 values of 53.39% and 53.39%, respectively, meaning that age and social economic status were likely to cause the heterogeneity.Significant heterogeneity introduced a warning about the generalization of the results. Thus, we performed subgroup analyses in order to find out probable causes for the significant heterogeneity. For EF and obesity risk, we found that the results of each subgroup were robust . All selected studies were adjusted for physical activity. As shown in I2 of each subgroup was higher than 50%. Therefore, it was difficult to identify the causes for the significant heterogeneity, which was possibly due to the different study designs among selected articles . We realized that the evidence level of cross-sectional studies was not as convinced as RCTs and cohort studies; however, considering the current research in EF, it is difficult to perform a meta-analysis on RCTs or cohort studies due to the lack of those studies at the time of our analysis. Third, although we performed subgroup analyses, we were unable to assess the exact effect of every exposure on the observed relationship due to limited selected studies. Third, inconsistent definition for EF should be taken into account. Also, the validity and reliability of food frequency questionnaire and dietary recall to evaluate EF and EI need to be considered. However, this is a common limitation in reviews and meta-analysis evaluating dietary behaviors , since tResults of the present meta-analysis suggest that increased EF is associated with lower obesity risk but higher EI. Despite the aforementioned limitations, these results suggest that increasing EF may benefit body weight management. However, other cofounding factors need to be considered as well, such as portion size. Thus, rigorously designed trials are warranted to confirm these results and to assess the practical applicability.


Conflicting results identifying the association between coiled-coil domain containing 62 (CCDC62) polymorphism, rs12817488, and Parkinson’s disease (PD) have been reported. To clarify whether rs12817488 is related to PD risk in Chinese population, we carried out this meta-analysis by searching literature from PubMed and Embase database regarding this polymorphism. Three eligible studies involving 1616 cases and 1649 controls were included in this meta-analysis. Our results showed statistically significant association between rs12817488 and PD risk in all four genetic models. Stratification by gender revealed similar results in both subgroup in these genetic models except for recessive model, in which rs12817488 was not significantly associated with PD in male subgroup. Unstable result was found in recessive model via sensitivity analysis, and publication bias was observed in recessive model as well, indicating that the pooled result from recessive model should be cautiously treated. Our meta-analysis implicates a possible relationship between rs12817488 and PD risk in Chinese population. Further validation of this association in large sample size study with different gender is warranted. Parkinson’s disease (PD) is the second most frequent neurodegenerative disease, clinically characterized with muscle rigidity, tremors, postural instability, bradykinesia and psychiatric symptoms, and costs on patients, caregivers, and society. The etiology of PD involves both genetic and environmental factors. Fewer than 5% of overall PD cases are attributed to genetic mutations mainly in α-synuclein12A number of genes associated with PD risk have been found in both familial and sporadic Patients via genome-wide association studies (GWAS) and high throughput genotyping methods like next generation sequencing, including SNCA, LRRK2, PINK1, SLC45A3, ACMSD, HLA, GBA, RIT2, and CCDC6246To clarify whether rs12817488 is related to PD risk in Chinese population, we performed this meta-analysis, aiming to identify the contribution of rs12817488 to PD pathogenesis and to illustrate possible reasons for these conflicting results.Eligible studies were systematically searched in PubMed and Embase databases up to Oct 30, 2015, with keywords including “Parkinson’s disease or PD” and “CCDC62 OR coiled-coil domain containing 62 OR rs12817488” and “polymorphism or mutation or variation or SNP”. We also manually examined reference lists for other relevant publications.Studies were chosen if they met the following criteria: (1) evaluating association between rs12817488 and PD in Chinese population; (2) a case-control study; (3) available phenotype and allele frequencies data. Reviews, abstracts from conferences, republished or duplicate studies, studies with insufficient information for data extraction were excluded.The following information was collected: (1) first author and publication year; (2) country and ethnicity; (3) sample size and sex ration; (4) phenotype distribution and minor allele frequency (MAF).P < 0.05. The pooled odds ratios (OR) and 95% confidence intervals (CI) were calculated by the Z test to evaluate this association under the allelic (A vs. G), dominant (AA + GA vs. GG), recessive (AA vs. GA + GG), and additive (AA vs. GG) genetic models. Heterogeneity among studies was tested using Q test and I2 statistic. If PQ > 0.10 or I2 < 50%, the pooled OR was estimated by fixed-effect model. Otherwise, random-effect model was applied. Sensitivity analysis was conducted by sequentially excluding each study to assess the stability of the results. Publication bias was assessed by Begg’s and Egger’s tests. P < 0.05 was considered significant for all tests.We used STATA software 11.0  for all statistical analyses. The departure of rs12817488 frequencies from expectation under Hardy-Weinberg equilibrium (HWE) was assessed by chi-square in control group, and it was considered to be disequilibrium if P > 0.05). The genotype distributions of all studies are summarized in A total of 36 studies were retrieved . As the result in Embase was the same with that in PubMed, therefore, 18 studies were retrieved. 1 review and 14 irrelevant studies were excluded. Finally, 3 eligible studies (1616 cases and 1649 controls) published from 2013 to 2015 were chosen, and the data were extracted67Overall, heterogeneity in the four genetic models was not statistically significant, and the ORs and 95% CIs were therefore calculated in fixed-effect model . Pooled P = 0.119; Stratification was performed by gender , and significant association was found in all genetic models within the female subgroup , Fig. 2.P = 0.027; OR = 1.46, 95% CI:1.19, 1.79, P = 0.000, respectively) in recessive model. In allelic, dominant, and additive models, the results were stable . The power were 0.992 and 0.991 respectively . In Liu RR’s study, patients with an age at onset <55 years were defined as EOPD, while others were described as LOPD. In Yu RL’s study, the cutoff age was 45 years of age. Li NN Some limitations existed in the current meta-analysis that must be considered. First, we only searched literature written in English language. Second, we only analyzed Chinese population. Third, we performed stratification only by gender, without referring other factors. Fourth, study and subject numbers were relatively small, and therefore, our result may be underpowered.Our meta-analysis suggests a role of rs12817488 in PD pathogenesis in Chinese population. However, future studies with even larger sample size different sex ratio, and different ethnic population should be done to confirm the findings of the current meta-analysis.How to cite this article: Lu, Y. et al. Possible association of CCDC62 rs12817488 polymorphism and Parkinson’s disease risk in Chinese population: a meta-analysis. Sci. Rep.6, 23991; doi: 10.1038/srep23991 (2016).


Purpose. The association between multiple sclerosis (MS) and fracture risk has been reported, but results of previous studies remain controversial and ambiguous. To assess the association between MS and fracture risk, a meta-analysis was performed. Method. Based on comprehensive searches of the PubMed, Embase, and Web of Science, we identified outcome data from all articles estimating the association between MS and fracture risk. The pooled risk ratios (RRs) with 95% confidence intervals (CIs) were calculated. Results. A significant association between MS and fracture risk was found. This result remained statistically significant when the adjusted RRs were combined. Subgroup analysis stratified by the site of fracture suggested significant associations between MS and tibia fracture risk, femur fracture risk, hip fracture risk, pelvis fracture risk, vertebrae fracture risk, and humerus fracture risk. In the subgroup analysis by gender, female MS patients had increased fracture risk. When stratified by history of drug use, use of antidepressants, hypnotics/anxiolytics, anticonvulsants, and glucocorticoids increased the risk of fracture risk in MS patients. Conclusions. This meta-analysis demonstrated that MS was significantly associated with fracture risk. Multiple sclerosis (MS) is a chronic inflammatory demyelinating disease of the central nervous system (CNS) affecting primarily young adults . MS is aRecently, several studies suggested that MS was associated with fracture risk –14. HoweWe searched the electronic databases PubMed, Embase, and Web of Science using the following search terms: (multiple sclerosis or MS) and  without restriction on language. The included articles were published before June 2014. All eligible studies were retrieved, and their references were examined manually for other potentially relevant studies. The inclusion criteria were as follows: (1) cohort design or case-control design; (2) the association of MS with risk of fracture that should be evaluated.All data were independently collected from the included studies according to a standardized protocol by two investigators. The discrepancies during data extraction were resolved by consensus. The same data in different studies were used only once. The following information was extracted: first author's name, publication year, study design, age, gender, follow-up years, sample size, and covariants.Z test was used to assess the pooled RR with the significance set at P < 0.05. The presence of between-study heterogeneity was evaluated by using the I2 statistic test, which does not inherently depend on the number of studies in the meta-analysis and is preferable to the test of heterogeneity. The value of I2 ranged from 0 to 100%. If obvious heterogeneity was observed among the studies (I2 > 50%), the random-effects model (the DerSimonian and Laird method) was used to calculate the pooled RR and 95% CI. Otherwise, the fixed-effects model (the Mantel-Haenszel method) was adopted for the meta-analysis. Subgroup analyses according to the site of fracture, gender, and history of drug use were also performed. We carried out the cumulative meta-analysis. Sensitivity analyses were conducted to evaluate the effect of individual study on pooled results and assess the stability of results. The potential publication bias was detected with Begg's funnel plot, and the funnel plot asymmetry was assessed by Egger's linear regression test [The association between MS and fracture risk was assessed using risk ratio (RR) with 95% confidence interval (CI). The ion test . All staA total of 9 cohort studies with 9,229,368 subjects met the inclusion criteria . The meaP < 0.00001). This result remained statistically significant when the adjusted RRs were combined . Subgroup analysis stratified by the site of fracture suggested significant associations between MS and tibia fracture risk , femur fracture risk , hip fracture risk , pelvis fracture risk , vertebrae fracture risk , and humerus fracture risk . However, there was no significant association between MS and ribs fracture risk  and radius/ulna fracture risk . In the subgroup analysis by gender, female MS patients had increased fracture risk . In contrast, no significant association was observed in the male MS patients . When stratified by history of drug use, use of antidepressants, hypnotics/anxiolytics, anticonvulsants, and glucocorticoids increased the risk of fracture risk in MS patients (As shown in patients .As shown in P = 0.22).Publication bias of the selected articles was assessed by Begg's funnel plot and Egger's test. The shape of the funnel plot did not show obvious publication bias . SimilarIn the current meta-analysis with 9,229,368 subjects, we found that there was a significant association between MS and fracture risk, even after excluding the studies without adjusting covariants. Further stratified analysis revealed that MS was significantly associated with fracture risk in female patients but not in male patients. This may be explained by lower bone mass in women with MS . After sThere were several possible reasons to explain the finds of this meta-analysis. First, MS patients usually had decreased bone mineral density (BMD). Dobson et al.  found thBetween-study heterogeneity is common and should be explored in the meta-analysis. In the current study, significant heterogeneity was found in the association of MS with fracture risk. Therefore, subgroup analyses were performed to explore the sources of between-study heterogeneity. The results indicated that gender might be the source of heterogeneity. In the male and female subgroups, the heterogeneity was decreased significantly. Sensitivity analysis revealed that the omission of any single study did not have significant impact on the overall meta-analysis estimate. Furthermore, in the meta-analysis, funnel plot did not reflect considerable asymmetry and Egger's test also indicated no obvious publication bias. All these made the meta-analysis results reliable to some extent.There are some limitations in this meta-analysis. First, the retrieved literature was potentially not comprehensive enough. We did not track the unpublished articles to obtain data for analysis. The potential effect of this publication bias was unknown. Second, fracture is a multifactorial disease and potential interactions among MS-environment should be considered. Moreover, as many other factors such as age may participate in the progression of fracture, we did not carry out subgroup analysis based on these factors due to limited data. Finally, the severity of MS could potentially be a confounder. Most studies included MS patients with mild to moderate severity, but more severe patients were the ones with walking issues and falling issues. However, we did not adjust the severity of MS in this meta-analysis. More studies should include the severity of MS, such as the Expanded Disability Status Scale (EDSS), as a confounder in the future.In conclusion, a significant association was detected between MS and fracture. Moreover, further studies with large sample size will be necessary to validate this result.


TPH1 (rs10488682) and progression of idiopathic scoliosis (IS) in Eastern European population sample. A total of 105 patients and 210 healthy gender-matched controls were enrolled in this study. The TPH1 promoter polymorphism was genotyped by amplification followed by restriction. The statistical analysis was performed by Fisher's Exact Test. The results indicated that the genotypes and alleles of TPH1 (rs10488682) are not correlated with curve severity, curve pattern, or bracing. Therefore, the examined polymorphic variant could not be considered as a genetic factor with modifying effect of IS. In conclusion, this case-control study revealed no statistically significant association between TPH1 (rs10488682) and progression of IS in Eastern European population sample. These preliminary results should be replicated in extended population studies including larger sample sizes. The identification of molecular markers for IS could be useful for a more accurate prognosis of the risk for a rapid progression of the curve. That would permit early stage treatment of the patient with the least invasive procedures.The concept of disease-modifier genes as an element of genetic heterogeneity has been widely accepted and reported. The aim of the current study is to investigate the association between the promoter polymorphism Idiopathic scoliosis (IS) is known for centuries but its etiology still remains unclear.The studies on IS genetics have indicated substantial heterogeneity in the etiology of the disease. First, there are predisposition genes that usually have low penetrance and are associated with a moderate increase of the risk for development of IS. In addition to predisposition to the development of deformity, genetic factors could also influence the curve progression . The conMelatonin is a focus of studies of the mechanism underlying the development of deformity and genes involved in melatonin signaling or biosynthesis are possible candidate-genes for IS.MTNR1A) were not associated with IS, in a linkage study of 47 American families [Genetic variants of melatonin receptor 1A (families  or Amerifamilies .MTNR1B) were not associated with IS, in Chinese females (473 cases/311 controls) [Three polymorphic variants in the coding region of melatonin receptors 1B (ontrols) . This asontrols)  and 406 ontrols) ), in Hunontrols) , in Bulgontrols) . N-acetyltransferase (AANAT) in a Chinese study (103 cases/107 controls) [GPR50) in the United States (406 cases/479 controls) [ O-methyltransferase (ASMT), previously known as hydroxyindole O-methyltransferase (HIOMT) and protein kinase C delta (PKCD) in a separate American cohort  [Some components of the melatonin pathway not associated with IS included aralkylamineontrols)  and in aontrols) , G prote TPH1 (rs10488682 T/A) was associated with the genetic predisposition to the adolescent idiopathic scoliosis (AIS) in Chinese cohort study (103 cases/107 controls) [ TPH1 gene are prone to be resistant to brace treatment [The promoter polymorphismontrols) , but thiontrols)  or Ameri1 rs1048882 T/A wareatment . MTNR1B (rs4753426) and IS predisposition or curve progression [ TPH1 (rs10488682) and progression of idiopathic scoliosis (IS) in Eastern European population sample.Our previous study in Bulgarian population found no association betweengression . The aimIn order to fulfill this aim, the following associations between the SNP and (i) curve severity (case-control study), (ii) curve pattern (case-only study), and (iii) brace treatment outcome (case-only study) were explored among Bulgarian patients.A total of 105 patients and 210 healthy gender-matched controls were enrolled in this study. All participants were included only after the subjects or parents of the minor patients signed a written informed consent.n = 86) and male (n = 19) patients were included.Patients with IS were recruited with the help of orthopaedic surgeons from Tokuda Hospital Sofia. Diagnosis was confirmed radiologically. The curves were measured by the Cobb method. Secondary scoliosis was excluded. The mean age at the beginning of the disease was 11.2 ± 3.1 years. Female (n = 84) and nonprogressive/slowly progressive scoliosis (n = 62) were compared to the control group. In this way the possible association between TPH1 and curve severity was examined.For the aims of the case-control study the patients were divided into two groups according to the preoperative or the last follow-up Cobb angle measurement: progressive scoliosis (mean Cobb angle = 62.7 ± 17.4°) and nonprogressive or slowly progressive scoliosis (mean Cobb angle = 22.1 ± 6.3°). Then these groups of progressive (n = 62), thoracolumbar (n = 31), and lumbar (n = 12) and then compared to each other. In this way the possible association between TPH1 and curve pattern was examined. Additionally, progressive cases (n = 84) were separated into two groups, with (n = 49) and without (n = 34) previous brace treatment, and compared to each other. In this way the possible association between TPH1 and brace treatment outcome was examined.For the aims of the case-only study all cases were divided into three groups according to curve pattern, thoracic  on automated high throughput nucleic acid isolation platform . TPH1 promoter polymorphism was genotyped by amplification followed by restriction.TheμL containing 100 ng DNA and 10x Prime Taq buffer , 10 mM dNTPs Mixture , 20 pmol forward and reverse primers , and 0.1 U Prime Taq DNA Polymerase . PCR amplification was performed in an AB 2720 Thermocycler  with an initial denaturation at 94°C for five minutes and a final extension of seven minutes at 72°C. The following thermal cycle was repeated 30 times: denaturation at 94°C for 30 seconds, annealing for 30 seconds at 58°C, and extension at 72°C for 30 seconds.The primer set consisting of forward primer 5′-AAGAAGTTGCACAATGCAGACA-3′ and reverse primer 5′-GTTGGGAAGACTGCAAGAAGC-3′ was used. The polymerase chain reaction (PCR) was carried out in a reaction mix of 20  SpeI , according to the manufacturer's instructions, and the restriction fragments were separated on agarose 3% gel in VG-SYS Horizontal Electrophoresis System . The lengths of the fragments representing the genotypes are 260 bp (AA), 146 + 114 (TT), and 260 + 146 + 114 bp (TA), respectively.The restriction fragment length polymorphism (RFLP) analysis was performed with the endonucleasep < 0.05 was considered to be statistically significant for comparison between data sets. Odds ratios (OR) were calculated with 95% confidence interval (95% CI). Statistical analysis was conducted with the IBM SPSS 19.0  software package for Windows.The statistical analysis was performed by Fisher's Exact Test to make genotype and allele comparisons between cases and controls as well as test for Hardy-Weinberg equilibrium. A value of  TPH1 (rs10488682 T/A) and IS progression were conducted in Eastern European population.A case-control study and a case-only study to investigate the association betweenCurve severity and curve pattern play a role in progression of IS. The general rules of progression are that larger curves carry a higher risk of progression than smaller curves, and thoracic and double primary curves carry a higher risk of progression than single lumbar or thoracolumbar curves . We sepa ERα and TPH-1 genes are independent factors predictive of bracing effectiveness [As a main nonoperative treatment for IS patients, brace treatment has proved to be effective in preventing curve progression by many studies conducted –18. The tiveness .We also divided the cases in subgroups according to brace treatment outcome and then compared the genotype and allele frequencies in the different subgroups of patients (case-only study) under genotypic  and allelic model. TPH1 promoter polymorphism were similar in the progressive group (Cobb angle >40°) compared to the control group (p > 0.05). In conclusion, the genotypes and alleles of TPH1 (rs10488682 T/A) could not be associated with the curve severity of IS among Bulgarian patients. This result corresponds to the previous American population study  [Genotypes were in Hardy-Weinberg equilibrium. The genotype and allele frequencies of the1 rs1048882 T/A coOdds ratios of genotypes and alleles in the subgroups are summarised in  TPH1 promoter polymorphism were comparable between the subgroups with different curve pattern (p > 0.05). On the basis of these results, the genotypes and alleles of TPH1 (rs10488682 T/A) could not be associated with the curve progression of IS among Bulgarian patients. A Chinese case-only study found no significant differences of the genotype or allele distribution between the groups of patients with different curve pattern [The genotype and allele frequencies of the pattern . The optOdds ratios of genotypes and alleles are summarised in  TPH1 promoter polymorphism were also comparable between the progressive cases with and without previous brace treatment (p > 0.05). In conclusion, TPH1 (rs10488682 T/A) could not be associated with the brace treatment final outcome. The observed differences in the results between our study in Eastern European population group and the study in Chinese population sample [The genotype and allele frequencies of then sample  could beOdds ratios of genotypes and alleles are summarised in  TPH1 (rs10488682) are not correlated with curve severity, pattern, or bracing. Therefore, the examined polymorphic variant could not be considered as a genetic factor with modifying effect of IS. These preliminary results should be replicated in extended population studies including larger sample sizes.The results indicated that the genotypes or alleles of TPH1 (rs10488682) and progression of IS in Eastern European population.In conclusion, this case-control study revealed no statistically significant association betweenThese results could not exclude a role of this polymorphic marker in the etiopathogenesis of IS in other population groups.The identification of molecular markers for IS could be useful for a more accurate prognosis of the risk for a rapid progression of the curve. That would permit early stage treatment of the patient with the least invasive procedures.


CRP gene believed to influence the protein levels have been extensively investigated in CAD community. Most of the published studies, however, have reported mixed findings. The objective of the present study was to examine the associations of CRP variants  with genetic risk of CAD by use of a meta-analysis.C-reactive protein (CRP) is a commonly used inflammatory marker and elevated CRP levels are shown to increase the risk of coronary artery disease (CAD). Sequence variations in the >C was not associated with CAD risk when all data were pooled together, nor did we find a significant association in subgroup analyses. Meta-analysis of +1444C>T studies showed a similar trend. However, a borderline association with CAD risk was revealed for −717A>G .The human case–control studies were identified through online search, hand search, and contacting the authors of original articles. We performed both random-effect and fixed-effect meta-analysis to estimate CAD risk . This analysis combined 16 studies in total. We found +942GCRP gene variants examined may not modulate CAD risk.These data suggest that the  Many groups have set out to characterize the pathological role of biomarkers related to inflammatory reactions in the initiation of CAD,5 in an attempt to expand the current knowledge on the heterogeneous etiology and molecular mechanism, thereby lowering the incidence of this aggressive disease.Coronary artery disease (CAD) is the foremost cause of preventable death across the globe. The number of deaths caused by heart disease has reached to 596,339 in the United States in 20117 Alternations in serum levels of CRP have biological significance including reflection of inflammation stages and higher CRP levels predispose individuals to various diseases. A series of clinical studies have shown that upregulated CRP serum levels result in increased likelihood to develop CAD.9 This observation was recently confirmed in multiple reports, where baseline CRP levels that may be affected by genetic factors were suggested to link with prognosis of cardiovascular diseases and short-term mortality.11 CRP levels vary substantially between individuals, which suggests an inheritable component in the regulation of serum levels. Available data have documented that an almost half of the variance could be attributable to genetic variants.13A most commonly used inflammatory marker is C-reactive protein (CRP), an acute phase reactant frequently presenting in atherosclerotic lesions.CRP gene on chromosome 1q21–23 contains 2 exons. Data on association between the variants at CRP locus and the coding of amino acids are currently limited and +942G>C (rs1800947) was previously reported to exert no influence on the amino acid.15 Several lines of evidence have established a significant connection of genetic variants with CRP levels, including +942G>C, −717A>G (rs2794521), and +1444C>T (rs1130864).17 The functional properties of these variants prompted a number of investigators to evaluate their contributions to genetic risk of CAD. The published studies, however, have produced mixed findings.20 More importantly, in a previous genetic analysis published in 2013, Li et al21 identified a significant increase in risk of CAD in relation to +942G>C. This effect estimation is possibly incorrect, because 2 papers with more than 2000 subjects were mistakenly categorized into “African” and “Asian,” respectively. Herein, we hospitalized that the aforementioned CRP gene variants may be genetic risk factors for CAD. To test the hypothesis, we chose to perform a meta-analysis.The human CRP variants and CAD risk. We first searched the Embase, PubMed, Web of Science, CNKI, and CBM databases. We used search terminology (polymorphism) AND (CRP) AND (CAD) and their synonyms  to identify studies. Then we carefully checked the studies analyzed in a meta-analysis for the association of CRP variants with CAD risk21 and hand searched the citations quoted in the single studies eligible for this meta-analysis. We also contacted the authors of a study with incomplete data in the original article,22 although no response was received. The study was approved by the ethics committee of General Hospital of Beijing Military Command of PLA.A 3-stage literature search was carried out to identify the studies of had a case–control design;CRP variants; andevaluated the association between genetic risk of CAD and at least one of the published genotype frequency in detail or the data could be obtained after contacting the author.We selected eligible trials without restraints on a minimum of sample size or language used in publication. Inclusion criteria required the trials:The studies satisfying all of the predescribed criteria were considered for further analysis. When 2 or more studies contained the same cases, we selected the largest study in which genotype frequency was detailed.P value for Hardy–Weinberg equilibrium (HWE). Disagreements were resolved via discussion.To maximize the accuracy of data, data abstraction was carried out independently by 3 investigators. Characteristics recorded for each trial included genotyped cases and controls, genotype distribution, first author, publication year, study country, racial descent (ethnicity), minor allele frequency in controls, and CRP variants. Calculation of summary ORs was carried out under the homozygous model, dominant model, and recessive model by use of a fixed-effect or random-effect meta-analysis. The extent of heterogeneity was measured by the χ2-based Q test. P values ≤0.05 were deemed statistically significant.23 The I2 metric was also used to test the proportion of total dissimilarity across the studies.24 I2 ≥ 50% indicated large heterogeneity. In a case of significant heterogeneity, the fixed-effect model was applied to summarize ORs for each trial25; otherwise, the random-effect model was performed.26 Stratified analyses by ethnicity, HWE (studies with or without HWE deviation), and sample size  were performed for CRP +942G>C variant. We used forest plots to show the combined results of all studies. HWE was tested for the included studies by a χ2 test. Publication bias was determined by 2 analytic tools-Begg's funnel plots and Egger's linear regression analysis.27 The leave-one-out sensitivity analysis was performed to check if the single studies had apparent influence on the overall meta-analysis results. The P values <0.05 were considered significant unless otherwise stated. All statistical data were analyzed using Stata software .We calculated odds ratios (ORs) along with 95% CIs to evaluate risk of CAD related to 38 As shown in Table >C, 5 for −717A>G, and 3 for +1444C>T. The Asian studies had a relatively smaller sample size compared with the Caucasian studies. Deviation from HWE was indicated in the genotype distribution of 4 studies for +942G>C.35 Moreover, there was wide difference in genotyping measurement and matching characteristics across studies.Study selection process is graphically represented in Figure >C did not have higher risk of CAD. Subgroup analysis by ethnicity, HWE deviation, and sample size demonstrated a similar nonsignificant trend toward increased risk of CAD. Figure h = 0.013, I2 = 51.6). With the aid of subgroup analysis and sensitivity analysis, we identified the study from the USA was the main cause of nonhomogeneous results.30 A significant increase in homogeneity was seen when this study was excluded . The overall effects were not influenced (data not shown).Table >G variant, we found a borderline reduction in CAD risk under the homozygous model   that may attribute to dissimilarities in study design and methodology   . Similarly, the tests presented no evidence of notable publication bias in the trials for −717A>G   and found no association with risk of CAD, and since then the association of common CRP variants with CAD risk has come under scrutiny. Replication efforts for the association with CAD include Xu et al19, who provided evidence supporting no contribution of +942G>C to CAD. Although this finding is consistent with the original study, it is challenged by a pilot case–control study, where the authors found +942C is associated with approximately 4-fold higher risk of CAD.20 For −717A>G CRP gene variant, published studies have shown the same contradictory results. According to Chen et al37, carriers of Chinese ancestry carrying −717A are genetically predisposed to CAD, which appears to contradict a study from Germany.31 In this work, Grammer et al analyzed all of the variants being investigated in this meta-analysis and detected no relationship between the genetic variants, previously shown to affect circulating CRP, and the prevalence of CAD. Several plausible reasons can explain the substantial inconsistency in these observations, including distinct genetic background, various errors in methodology, and nonstandard selection of control subjects. For example, Grammer et al, unlike the other investigators who employed healthy individuals as controls, considered patients with stenoses <20% as controls; this classification may lead to imprecise assessment of the association.31Since sequence variants of the CRP variants and CAD risk by means of meta-analysis. A total of 16 articles comprised of 14 studies for +942G>C, 5 for −717A>G and 3 for +1444C>T were summarized in this analysis. For +942G>C CRP gene variant, we performed both global analysis and stratified analyses, with none of the analyses indicating a significant genetic association. This lack of an association with CAD persisted when data for −717A>G and +1444C>T were analyzed. These findings fail to confirm the previous epidemiological and meta-analysis data supporting a strong association between the CRP variants and incident CAD. Li et al21 analyzed 6751 Caucasians and found increased CAD risk associated with +942G>C genotypes. Conversely, we failed to replicate this significant association by use of a meta-analysis combining 9186 Caucasians. The null association indicated in the present study highlights the importance of a sufficient sample size to derive a precise evaluation of a genetic association. Therefore, whether the CRP gene variants act as modifiers of CAD risk remains to be validated in a large-scale study.In view of these mixed findings, we were spurred on to evaluating the relationship between 46 targeted the PlA1/A2 variant of GPIIIa and found that the PlA2 allele is a risk factor of ischemic stroke. In a recent study, presence of the same variant was reported to increase the risk of stroke and MI.47 Many other candidate gene polymoprhisms, such as CaMK4 and GRKs, are frequently shown to induce hypertension.49 These findings suggest that variants are likely to play a pivotal role in the development of cardiovascular disorders. The null results shown in the present work need to be verified in a new study, possibly in a sufficient number of subjects.The role of genetic variants in the carcinogenesis of cardiovascular disease has received widespread attention in recent years. For example, Lanni et al>C and −717A>G. We identified the major sources of heterogeneous results, and noted that the combined effects remained stable when the outlawed study was excluded, suggesting the impact of heterogeneity on overall estimates exists but not substantial. Second, many previous studies support an association of CRP variants with incident CAD, which nevertheless is not confirmed in the present study. It is possible that CRP variants are low-penetrance polymorphisms and the minor effects on common diseases require a sufficiently large study to detect. Third, several studies of +942G>C deviated from HWE, and it is the deviation that may affect the precision of overall meta-analysis results. Fourth, CAD is genetically heterogeneous. In addition to inherited genetic factors, lifestyles and exposure to environmental risk factors are important components in the malignant progression. Further research is clearly required to investigate the pathogenesis of this common cardiovascular disorder. Finally, this article only focused on the gene effect on susceptibility to CAD. The effect of the disease modifying  or responsibility to the medication was not studied.In this analysis, multiple factors may limit the extent the results can be generalized to. First, significant heterogeneity was indicated across the studies for +942G>C and CAD risk in Caucasians, and lack of a relation for −717A>G and +1444C>T. In addition, to our knowledge, this is the first study addressing the relationship between all commonly studied CRP variants and CAD risk, facilitating a more precise evaluation and a better understanding of the role of CRP in CAD.This analysis also has several strengths. The first strength refers to the novel findings not discovered in earlier quantitative assessment, including the null association of +942GCRP genetic variants previously reported to influence circulating C-reactive protein may not be linked to the development of CAD. This lack of association requires future larger studies to confirm or refute.In conclusion, this study suggested that the