NRAMP1) and the risk of pulmonary tuberculosis (PTB), Revealing inconclusive results. This study aimed to investigate the correlation between the NRAMP1 3’UTR polymorphism and the risk of PTB.Many studies have investigated the association between the 3’UTR polymorphism in natural resistance-associated macrophage protein 1  and 95% confidence interval (95% CIs) were calculated to assess the strength of the association.NRAMP1 3’UTR polymorphism displayed a significant positive correlation with the risk of PTB in 3 models . A stratified analysis by ethnicity revealed that the NRAMP1 3’UTR polymorphism was associated with an increased risk of PTB in the Asian population, but not in Caucasian, African, and South American populations.These 29 case–control studies included 4672 cases and 6177 controls. The NRAMP1 3’UTR polymorphism may be considered a risk factor for PTB in the Asian population.The present results indicate that the  In 2016, 10.4 million new cases were reported, and 1.7 million individuals died of TB. Most of the individuals with TB reportedly reside in India China, Philippines, Pakistan, Nigeria, and South Africa. The precise mechanisms underlying TB pathogenesis remain unknown. Weakened immunity, human immunodeficiency virus (HIV) infections, alcohol abuse, advanced age, chronic corticosteroid therapy, diabetes, socio-economic status, and malnutrition are the prominent risk factors of TB. Not all individuals exposed to the similar risk factors develop tuberculosis, thus suggesting the involvement of genetic factors, including single-nucleotide polymorphisms, in TB pathogenesis.Tuberculosis (TB), a chronic infectious disease caused by M tuberculosis infects the body, the immune system produces a barrage of defensive protein molecules, recruiting phagocytic cells, which eliminate the pathogen. Natural resistance-associated macrophage protein 1 (NRAMP1) is also referred to as the solute carrier family 11 proton-coupled divalent metal ion transporter membrane1 (SLC11A1), which plays an important role in the immune response to mycobacterial infections. NRAMP1 encodes a divalent transition metal (Fe and Mn) transporter that localizes on the lysosomal membrane. Iron is an essential mycobacterial nutrient that also influences the generation of reactive oxygen and nitrogen intermediates. Therefore, NRAMP1 is involved in resistance to intracellular pathogens, including Leishmania, Salmonella, and Mycobacteria. NRAMP1 is associated with various infectious diseases and inflammatory diseases. It primarily contains 4 polymorphisms, rs17235416 (3’UTR), rs17235409 (D543N), rs3731865 (INT4), and rs34448891 (5=(GT)n). Among these polymorphisms, 3’UTR polymorphisms have been widely investigated for their association with TB. A functionally significant TGTG del allele in NRAMP1 leads to reduced production of NRAMP1 when compared to the TGTG+ allele and may be correlated with the risk of TB.When NRAMP1 3’UTR polymorphism and the risk of pulmonary tuberculosis (PTB) in different regions,–29 revealing inconclusive results. A case–control study by Medapati et al reported that the NRAMP1 3’UTR polymorphism is significantly associated with the susceptibility to TB among Indian individuals. However, Jafari reported no significant association between these 2 aspects. Hence, the present meta-analysis aimed to better and comprehensively assess the correlation between the NRAMP1 3’UTR polymorphism and the risk of PTB.Numerous studies have investigated the association between the 22.1We performed a comprehensive search in PubMed, Elsevier, and the Cochrane Library , using the following keywords Ethical approval was not necessary since this study is a meta-analysis.(“tuberculosis” or “TB” or “mycobacteria”), (“polymorphism” or “mutation” or “variant”), and . Articles in all languages were included. Furthermore, additional related articles were identified through screening of the references of relevant articles. In case of a duplicate publication, the largest study was selected. The present study was conducted in accordance with the PRISMA guidelines for systematic reviews and meta-analyses.2.2Following were the inclusion criteria of this meta-analysis:(1)NRAMP1 3’UTR polymorphism and the risk of PTB;case–control or cohort studies on the (2)sufficient data estimating odds ratios (ORs) with 95% confidence intervals (95% CIs).The major exclusion criteria were as follows:(1)meta-analyses, letters, reviews, or editorial comments;(2)studies were not irrelevant to TGTG ins/del;(3)not PTB;(4)studies with insufficient data;(5)other than case–control study;(6)genotype distributions were not in accordance with Hardy–Weinberg equilibrium (HWE) in the control group.2.3 The NOS contains 3 categories which are selection(0–4 points), comparability(0–2 points), and exposure (0–3 points). The total scores ranged from 0 to 9.For all studies, EZ and LZ extracted the data independently and reached a consensus. Any disagreement was resolved through discussion among the 3 authors. The following information was collected from all eligible studies: the first author, year of publication, country of origin, ethnicity, source of controls, number of case and control subjects, HIV status, genotype frequencies in case and control groups, and HWE. We evaluated the quality of all of the studies included according to Newcastle–Ottawa scale (NOS).2.4NRAMP1 3’UTR polymorphism and the risk of PTB was evaluated using crude ORs with 95% CIs. To assess heterogeneity, the Cochrane Q test and the I-squared (I2) metric were used. If the I2 metric was <50%, indicating a lack of heterogeneity among studies, A fixed-effects model (the Mantel–Haenszel method) was used. Otherwise, the random-effects model (the DerSimonian and Laird method) was applied. In addition, subgroup analyses based on ethnicity  were carried out. Funnel plot analysis and Egger test were conducted to identify a publication bias of the meta-analysis. All statistical analyses were performed using STATA 11.0 software  and P values were 2-tailed.The association between the 33.1In accordance with the flowchart shown in Figure 3.2NRAMP1 3’UTR polymorphism was significantly associated with the risk of PTB in 3 models ; however, the recessive model did not reveal this significant association . In a stratified analysis by ethnicity, the present meta-analysis revealed that the NRAMP1 3’UTR polymorphism was associated with an increased risk of PTB in the Asian population , but not in the HIV+ individuals.Table 3.3t = 0.14, P = .888) and in the dominant model ; however, publication bias was detected in the del/del vs. ins/ins models .Funnel plot analysis and Egger's test were conducted to assess publication bias. Egger test did not reveal any evidence of publication bias in the Ins/del versus ins/ins models  scanning entire genomes for genetic variation include immense amounts of SNPs and have been designed for the same ethnic background and study design. A GWAS by Zheng et al reported that 2 loci 14q24.3  and 20p13 were significantly associated with TB in Han Chinese individuals. Miao et al reported that rs9272461 is significantly associated with the risk of PTB in various genetic models. Hence, GWAS are required to further validate the present results.Despite the strengths of the present study, some limitations should be acknowledged. First, upon subgroup analysis by ethnicity, the NRAMP1 3’UTR polymorphism may be associated with an increased risk of PTB in the Asian population, but not in the Caucasian, African, and South American populations. Well-designed epidemiological studies with larger sample sizes are needed to verify the present findings.In conclusion, the present results indicate that the www.editage.cn] for English language editing.We would like to thank Editage [Data extraction: Erjiang Zhao, Lin ZhuFormal analysis: Erjiang Zhao, Danning ZhangInvestigation: Yang Liu, Lin Zhu, Dan Ning Zhang.Methodology: Yang Liu, Erjiang Zhao, Dan Ning Zhang, Zhe Wang, Danning Ding.Resources: Yang Liu.Software: Erjiang Zhao, Lin Zhu.Writing – original draft: Yang Liu, Erjiang Zhao.Writing – review & editing: Yang Liu.Yang Liu orcid: 0000-0002-3877-9203.


Clusterin (CLU) is considered one of the most important roles for pathogenesis of Alzheimer’s Disease (AD). The early genome-wide association studies (GWAS) identified the CLU rs11136000 polymorphism is significantly associated with AD in Caucasian. However, the subsequent studies are unable to replicate these findings in different populations. Although two independent meta-analyses show evidence to support significant association in Asian and Caucasian populations by integrating the data from 18 and 25 related GWAS studies, respectively, many of the following 18 studies also reported the inconsistent results. Moreover, there are six missed and a misclassified GWAS studies in the two meta-analyses. Therefore, we suspected that the small-scale and incompletion or heterogeneity of the samples maybe lead to different results of these studies. In this study, large-scale samples from 50 related GWAS studies  were selected afresh from seven authoritative sources to reevaluate the effect of rs11136000 polymorphism to AD risk. Similarly, we identified that the minor allele variant of rs11136000 significantly decrease AD risk in Caucasian ethnicity using the allele, dominant and recessive model. Different from the results of the previous studies, however, the results showed a negligible or no association in Asian and Chinese populations. Collectively, our analysis suggests that, for Asian and Chinese populations, the variant of rs11136000 may be irrelevant to AD risk. We believe that these findings can help to improve the understanding of the AD’s pathogenesis. It is characterized by accumulation and toxic effect of the amyloid β-peptide (Aβ) deposits and neurofibrillary tangles in brain3. Previous studies predict that the newly diagnosed AD patients are expected to reach as many as 135 million by 2050 from about 35 million in 2009 around the world if lack of the effective preventive measures5.Alzheimer’s Disease (AD) is a commonest kind of neurodegenerative disorders with a complex pathogenesis, and has become one of the leading causes of death in elderly people8, and some of the variants at CLU can affect its expression level in brain10. Two early genome-wide association studies (GWAS) identified a single nucleotide polymorphism (SNP) rs11136000 (T < C) significantly associated with AD in the CLU gene by analyzing the large-scale Caucasian populations12. In particular, Harold et al.11 and Lambert et al.12 analyzed 11,756 and 14,490 individuals from USA, UK, Ireland, Germany, France, Italy, Spain, Belgium and Finland, respectively, and both of them found that the minor allele variant of rs11136000 can reduce the risk of AD  of odds ratio (OR) less than the value 1).Clusterin (CLU) is considered one of the most important roles for pathogenesis of AD by influencing the structure and neurotoxic effects of Aβ deposits18 and inconsistent28 results involved in Caucasian, Asian and African populations. For example, by analyzing 268 AD cases and 389 controls from China, Lin et al. find that the participants carrying 2 copies of minor allele in rs11136000 are associated with a decreased risk of AD17. The consistent result in North American Caucasian population is also identified by Carrasquillo et al.18. While in Canadian and Korean populations, the rs11136000 is found not associated with AD according to the studies of Bettens et al.24 and Chung et al.27, respectively. Then, two independent meta-analysis studies re-assess the results of these GWAS studies published before June 20, 2013 (18 studies) and August 31, 2014 (25 studies), respectively, and both of them found this SNP is significantly associated with AD in populations of Asian and Caucasian30. But among the subsequent 18 GWAS studies published after August 31, 2014, many of them report inconsistent results in the corresponding populations47. Moreover, by comparing the selected GWAS articles published before June 20, 2013 in the two meta-analysis studies, we find the selection is incomplete for both of them. In particular, Liu et al.29 miss two GWAS articles about Caucasian populations24, and Du et al.30 miss a GWAS article about Asian population27. In fact, through our further investigation, a total five related GWAS articles published before August 31, 2014 are not collected in the two meta-analysis studies52. In addition, a GWAS study about American and German populations is misclassified to the Asian ethnicity subgroup in Du et al.’s study22.However, the subsequent studies report consistent30) by searching the PubMed, ClinicalKey, AlzGene, Google Scholar, CNKI, Wanfang and VIP databases, and reevaluated the association between AD and rs11136000 polymorphism in Caucasian, Asian and Chinese population using the method of meta-analysis as previously described63. The use of more complete and larger scale samples would make the results more reliable.We suspected that the small-scale and incompletion or heterogeneity of the samples maybe lead to different results of these studies. In this study, we selected 50 related GWAS studies with large-scale samples from 40 articles , Wanfang (http://www.wanfangdata.com.cn/), CNKI (http://www.cnki.net/) and VIP (http://www.cqvip.com/) using the keywords: “Alzheimer’s disease”, “rs11136000”, “Clusterin” or “CLU”. The CNKI, Wanfang and VIP are very authoritative and reliable Chinese database. And then, we consulted the related studies collected in AlzGene database  which was a publicly available resource providing the information of AD genetic variants from 1,395 GWAS studies 64. In addition, we further queried references of these identified GWAS studies in previous steps and the articles citing them using the Google Scholar (http://scholar.google.com/).All of the possible studies were selected by searching the databases of PubMed  The study is a GWAS to analysis the association of rs11136000 polymorphism and AD. (2) It is a case-control design study. (3) The study provides both of the numbers of cases and controls. (4) The study provides the information about the ethnicity of each individual. (5) The detailed data for rs11136000 genotypes are available in the study.OR value and its 95% CI, as well as the corresponding P value). Moreover, if these results are not provided in the study directly, we would calculate them by the genotype data using the R program (http://www.r-project.org/).We extracted the related data for subsequent analysis from these identified studies: (1) each study’s publication date. (2) The first the author’s name in each of these studies. (3) The numbers of AD patients and controls of each study. (4) The sample’s ethnicity of each study. (5) The detailed genotype data of rs11136000 polymorphism both in AD patients and controls. (6) The types of genotyping platforms. (7) The key results of each study (i.e. the The rs11136000 polymorphism contains two types of variants (T and C). T is the minor allele and C is the major allele. We assumed that they are the lower and high risk factor for AD, respectively. Then, the dominant model , allele model (T versus C) and recessive model (TT versus TC + CC) were used in this study. According to Table P < 0.01 as previously described65. Briefly, for the SNP in each case and control group, the simulated P values were calculated to measure the deviation from HWE based on 10,000 iterations. The R package ‘Genetics’ was used to perform the HWE test (https://cran.r-project.org/web/packages/genetics/index.html).The HWE test of the rs11136000 polymorphism in AD patient and control groups was performed using a non-continuity correction chi-squared method with the significance level I2 value and Cochran’s Q. I2 value range from 0 to 100%, and it is calculated by Cochran’s Q according to the formula Q is based on a chi-squared distribution with k − 1 degrees of freedom, and k means the number of studies. Usually, the extreme, high, moderate and low heterogeneity was considered corresponding to the I 2 value of >75%, 50–75%, 25–50%, and <25%, respectively. In this study, the threshold of significant heterogeneity was set as I2 > 50% and P < 0.01 according to previous studies56.In this study, the heterogeneity among the kinds of populations was measured by the two parameters, 66. We used the R package ‘meta’ to perform the meta-analysis, and determine the significance level of association between rs11136000 and AD through the pooled OR value and its 95% CI, as well as the corresponding P value (http://cran.r-project.org/web/packages/meta/index.html). And then, the original samples were further split into Caucasian, Asian, East Asian and Chinese populations, and the meta-analysis was performed in these subgroups.According to the results of heterogeneity test, the random and the fixed effect model were performed when the heterogeneity was significant or not, respectively67 and Egger’s test68. The threshold of significant publication bias was set as P < 0.05. Then, we used the asymmetry of the funnel plots to describ the results of the publication bias analysis. Finally, for sensitivity analyses, we excluded each study in turn from the whole sample to measure the influence of each study.We first evaluated the publication bias of the studies used in dominant, allele and recessive model, respectively, by the two common checking methods, the Begg’s testAll the datasets used in this are available from the corresponding author.By a keyword search in the publicly available databases and a screening according to the criteria, a total 46 studies from 36 articles were identified which mainly involved in Caucasian and Asian populations. Moreover, a study about Sweden population was selected from AlzGene database, and three studies involved in Asian populations were identified by the citation check using Google Scholar.OR value and its 95% CI, as well as the corresponding P value were shown in Supplementary Table Figure P value of HWE to assess the genotype distribution of rs11136000 polymorphism in AD patients and controls separately. Using a significance level of P < 0.01, we observed that a few of the samples deviated from HWE, including the case samples from the study of Yu et al. (P = 9.0 × 10−3) and Gu et al. (P = 2.0 × 10−4), and the control samples from the study of Rezazadeh et al. (P = 1.0 × 10−4), Gu et al. (P = 1.0 × 10−4) and Lin et al. (P = 9.0 × 10−3). More detailed information about the results of the HWE test was described in Supplementary Table We calculated the I2 = 0% and P = 0.60), allele (I2 = 10% and P = 0.28) and recessive model (I2 = 33% and P = 0.04). Therefore, the meta-analysis with fixed effect model was performed to assess the association between rs11136000 and the risk of AD, and we found significant results in all the three models. In particular, the significant association between the minor allele (T) of rs11136000 and a decreased risk of AD was identified in the allele   and recessive model  , dominant  and recessive model      Fig. , but not49) Fig.  and rece26) Fig. .Figure 3OR = 0.939, 95% CI = 0.878–1.004, P = 0.0654)    Fig. , dominan70) Fig.  and rece41) Fig. , which wMoreover, given that a few samples from four GWAS studies (three Asian populations and a mixed population) deviated from HWE, we further tested whether they affected the accuracy of the results by removing these studies from whole sample, Asian, East Asian and Chinese subgroups, respectively. The results were consistent with what we had been observed previously in whole sample and the subgroups using allele, dominant and recessive model. Table P value of Begg’s and Egger’s test is 0.80 and 0.24, respectively, for dominant model. Similarly, the P value is 0.43 (Begg’s test) and 0.21 (Egger’s test) for the allele model, and 0.22 (Begg’s test) and 0.61 (Egger’s test) for the recessive model. Moreover, through the sensitivity analysis, for all the three genetic models, we did not found a significant change of the association level between rs11136000 and AD when excluding any of the studies. Supplementary Tables As the funnel plots show Fig. , we did 3, and previous studies reported that the CLU could markedly influence the fibrillary Aβ formation and accumulation to mediate its toxicity in vivo, and likely as one of the most important roles for pathogenesis of AD7. Then, the subsequent GWAS studies found some variants in CLU were differently distributed between AD patients and controls18. Among these variants, a significant association was found between the minor allele (T) of rs11136000 and a decreased risk of AD by Harold et al.11, Lambert et al.12, Carrasquillo et al.13 and Seshadri et al.14. However, these results could not be repeated in other populations by the following studies28.AD was characterized by accumulation and toxic effect of the Aβ deposits in brain30, many of the following studies also reported the inconsistent results47. Moreover, according to our further investigation, the two meta-analyses missed out a total six related GWAS studies published before August 31, 201452, and a GWAS study about American and German populations is misclassified to the Asian ethnicity subgroup in Du et al.’s meta-analysis22. Therefore, we suspected that the small-scale and incompletion or heterogeneity of the samples maybe lead to different results of these studies.Although the two independent meta-analyses found a significant association between the minor allele (T) of rs11136000 and a decreased risk of AD in Caucasian and Asian ethnicities by integrating the data from related GWAS studies published before June 20, 2013 (18 studies) and August 31, 2014 (25 studies), respectivelyOR = 0.829, 95% CI = 0.796–0.864, P < 0.0001), allele  and recessive model . Different from the results of the previous studies, however, rs11136000 polymorphism was found not associated with the risk of AD in Asian ethnicity using the dominant  and recessive model , as well as in Chinese population using the dominant , allele  and recessive model .In this study, 50 related GWAS studies (including the 6 missing and 18 novel studies) were selected afresh from seven authoritative sources, and the association level between rs11136000 and risk of AD in Caucasian, Asian and Chinese ethnicity was re-evaluated. We also found a significant association between rs11136000 polymorphism and the decreased risk of AD in Caucasian ethnicity using the dominant (As far as we know, our meta-analysis about the association of the CLU rs11136000 polymorphism with the risk of AD is by far the largest scale study. The results reveal a significant association between them in Caucasian ethnicity but not Chinese ethnicity, which is consistent with the findings of most of the corresponding GWAS studies. In summary, we believe that these findings can help to improve the understanding of the AD’s pathogenesis.Supporting Information


IL-1B, IL-10 and TNF-α. This study emphasizes the need for larger randomized controlled clinical trials to inform clinicians and patients about the role of genetic factors on dental implant survival and the success rate in healthy and compromised patients.Dental implants provide a predictable treatment option for partial and complete edentulism via the placement of a fixed permanent artificial root to support prosthetic dental crowns. Despite the high survival rates, long-term dental implant failures are still reported, leading to implant removals and additional financial and health burdens. While extrinsic factors that improve the success rate of implants have been well explored, the impact of genetic factors on this matter is poorly understood. A systematic review and meta-analysis study was conducted to determine whether genetic factors contribute to an increased risk of dental implant failures. A comprehensive search for peer-reviewed articles on dental implants and genetic factors was performed using various literature database libraries. The study design was conducted according to Preferred Reporting Items for Systematic Reviews and Meta-Analysis (PRISMA) guidelines, and the obtained records were registered in the International Prospective Register of Systematic Reviews (PROSPERO) database. According to the exclusion/inclusion criteria, 13 studies were eligible for this study out of 809 articles. The meta-analysis of the combined association studies of DNA variations and dental implants did not indicate an increased risk for implant failure due to DNA variations in  Association studies between common DNA variations and human diseases have been proven very useful for identifying genetic factors that increase the risk or provide protection to human complex diseases ,2. DNA vFailure of osseointegration occurs due to multifactorial conditions, including individual susceptibility or risk factors. Failure can occur even under proper conditions of bone tissue due to possible host immune responses ,14,15. CIL-2 and IL-6 genes were associated with an increase in the expression of both cytokines and with the development of chronic diseases such as periodontitis ) AND (“bone and bones” [MeSH Terms] OR (“bone” [All Fields] AND “bones” [All Fields]) OR “bone and bones” [All Fields] OR “bone” [All Fields]) AND (“genes” [MeSH Terms] OR “genes” [All Fields]); (“genetic therapy” [MeSH Terms] OR (“genetic” [All Fields] AND “therapy” [All Fields]) OR “genetic therapy” [All Fields] OR “genetic” [All Fields]) AND (“risk factors” [MeSH Terms] OR (“risk” [All Fields] AND “factors” [All Fields]) OR “risk factors” [All Fields]) AND  OR “dental implants” [All Fields]) AND failure [All Fields];  OR “dental implants” [All Fields]) AND failure [All Fields] AND (“genetic therapy” [MeSH Terms] OR (“genetic” [All Fields] AND “therapy” [All Fields]) OR “genetic therapy” [All Fields] OR “genetic” [All Fields]) AND (“risk factors” [MeSH Terms] OR (“risk” [All Fields] AND “factors” [All Fields]) OR “risk factors” [All Fields]).Implantology, Clinical Implant Dentistry and Related Research, Clinical Oral Implants Research, European Journal of Oral Implantology, Implant Dentistry, International Journal of Oral and Maxillofacial Implants, International Journal of Oral and Maxillofacial Surgery, International Journal of Periodontics and Restorative Dentistry, International Journal of Prosthodontics, Journal of Clinical Periodontology, Journal of Dental Research, Journal of Oral Implantology, Journal of Oral and Maxillofacial Surgery, Journal of Oral Rehabilitation, Journal of Periodontal Research, Journal of Periodontology, and Journal of Prosthetic Dentistry. Additionally, a manual search was conducted in the references of included articles.A manual search from September 2017 until February 2018 was conducted for the following journals: Three previously calibrated reviewers  selected the articles and performed the data collection. Discrepancies in the analysis were solved in a consensus meeting for analysis of the selected titles and abstracts, with an agreement test value for the selected articles in the three databases. In order to decrease bias in the selection of the articles, authors G.A.H.K. and C.B. participated in the selection of the sample, data collection and examination of the databases as well. Consensus meetings for the selection of each article within the selected sample pool were held on weekly bases . The group and other researchers worked together to consolidate the analysis of the topics (P.P.S. and W.D.F.).Data extracted from studies that passed the inclusion criteria were analyzed and the main standardized information were obtained as follows: (1) author; (2) year of publication; (3) country and origin of the study; (4) number of patients; (5) analyzed group of patients; (6) mean age of the patients; (7) number and sites of implants; (8) trade mark of the implants; (9) implants failure; (10) main data of the implants; (11) data of periodontal evaluation; (12) studied variables; (13) methodologies; (14) rate of peri-implant bone loss; (15) follow-up period of each study. Data were collected using a standardized file built in Excel software.The selected clinical studies were evaluated based on the consideration of their methodology approach on a bias scale by the National Health and Medical Research Council . The bias scale allows for evidence levels of different categorical studies . Therefore, the NHMRC establishes a hierarchy of studies, classifying clinical studies at different levels. For intervention studies: I-systematic review; II-Randomized controlled study; III-1 pseudorandomised controlled study; III-2 clinical study with a group control; III-3 a comparative study without control group; IV case series ,29,30. p < 0.05. Reviewer Manager 5.3 (Cochrane Group) was used for the meta-analysis and graphic elaboration. The quantitative data collected from the articles was tabulated for the analysis of odds ratio (OR) with a correspondent 95% confidence interval (CI). For all analysis, significant values were considered as The primary outcome was to analyze if there is a significant association between genetic factors and the failure of dental implants based on previously published studies that passed the inclusion criteria.p < 0.1 and was evaluated using the Q (x2) test and I2 value. Statistical I2 value was used in the analysis of heterogeneity variations, and values above 75 (0–100) were considered to indicate significant heterogeneity [A fixed-effects model was applied in case no significant differences were observed in the data. Alternatively, a random effects model was applied in case significant differences (high heterogeneity among the tests) were observed. Heterogeneity was considered significant at ogeneity ,31,32,33ILs and TNFα were made in order to avoid potential heterogeneity due to different groups of patients (failure vs. control group) [Sensitivity tests for the analysis of patient subgroups and the allele frequency of different types of alleles for the genes l group) ,24. A total of 809 articles were identified from the three online databases searches. After applying the inclusion/exclusion criteria, 50 articles were eligible for full-text assessment. After a complete assessment, 13 articles fulfilled the inclusion criteria for qualitative analysis and 6 studies for meta-analysis, as shown in Eleven out of the 13 selected studies were from South America (Brazil) ,21,33,34IL-1B [IL1A and IL1B [IL1B and IL1RN [IL-2 [IL-4 [IL-10 [TNFα [MMP8 [GFβ1 [RANKL [For statistical analysis different kinds of cellular mediators were evaluated based on PICO criteria, with a focus on selecting studies that provided data on genetic factors that can lead to implant failure, such as vitamin D receptor polymorphism D , interleIL-1B , IL1A anand IL1B ,35, IL1Bnd IL1RN , IL-2 [1-2 [IL-4  and IL-14 [IL-10 ,15,34, T10 [TNFα , metallo1 [RANKL . The sel1 [RANKL ,20,21,341 [RANKL ,15. One 1 [RANKL , and ano1 [RANKL .MMP-8 presented significant effect for T allele in 76.25% of the study group (failed implants). The genotype T/T in 63.75% of the study group was indicative of early loss of the implants (p = 0.0011). The C/T genotype was found in 48% in the control group (with no implant failure), while in 63.75% of the patients was observed T/T genotype (p = 0.0009) [IL4 gene, SNP-590, the C allele was associated with implant loss  for [IL-1A and IL-1B alterations and found that alleles 1 and 2 of both cytokines could be associated with success or failure of the dental implants. They also addressed environmental factors such as smoking and alcohol use and showed that there was no significant association with implant failures. The other studies did not indicate additional genetic risk factors for the failure of dental implants, as shown in In relation to the impact of the genetic alterations on dental implants survival, only one study about the role of 2.4) for . Finally2.4) for  analyzedA total of 2130 patients combined from 13 studies were included for this systematic review and meta-analysis. Among them, only one study considered a single group of patients without including a control group . The othp = 0.003), posterior/anterior (p = 0.037), mean length of the implant (p = 0.001), primary stability (p = 0.001), surgical technique (p = 0.016), the quality of bone tissue (p = 0.049), and edentulism (p = 0.009) on dental implant survival. Dirschnabel et al. (2011) [p = 0.019), site of implant installation (p = 0.001) and medical/systemic conditions of the patient (p = 0.04). Similarly, Pigossi et al. (2014) [p = 0.031), maxillae/mandible (p = 0.003), anterior/posterior regions (p = 0.037), primary stability  as factors relating to implant survival. Montes et al. (2009) [IL1RN (intron 2) was associated with failure of implants in individuals with multiple dental implant losses in addition to factors such as edentulism and the number of teeth present. In addition, some studies analyzed the periodontal condition of the remaining teeth, detecting significant differences in probing depth index in comparison to the control groups,   identifie p = 0.0, the qua p = 0.01, the qua14  identI2 > 75%, p < 0.1) was not observed in this study. Acceptable measures (I2) were identified for study group of IL10 (Failure group), IL1B (Control G vs. Failure G), IL10 (Control G vs. Failure G), TNF-α (Control G vs. Failure G).Studies that analyzed the same allele of the same gene of interest were grouped together to increase the power of statistical analysis and the level of significance by comparing the control groups with no dental implant failure (control group) and the group with implant failure (study group). An increase of power analysis can be achieved by having a larger sample size of experimental and control groups with similar environmental conditions. Therefore, this study sought to combine all related studies that analyzed the association between DNA variations of genetic factors and dental implant failure. For each DNA allele, two studies were combined for the frequency distribution of genetic alleles. A high heterogeneity in the frequencies of the alleles and OR across the individual studies  or allele (IL1B-C511) of IL1B. The genetic data was divided into four subgroups including the prevalence of IL1B-C511 allele in study (failure) group and control group, and the prevalence of the IL1B-T511 allele in the failure group and control group. The subgroups for each allele were combined from the two individual studies and analyzed for an association between study groups from both studies and dental implant failure [IL1B-C511 in the study group (failure G) and control group (control G) shows a pooled OR of 0.85, 95% CI of 0.62–1.18, and p value 0.33, indicating no significant association in carrying this allele and increase risk for implant failure , p = 0.33 , , . As wellp = 0.92 .The homogeneity analysis for the comparison between study groups (failure) vs. the control group was performed among the selected publications in this study. The data shows low levels of heterogeneity for all the alleles included in the analysis A–F.Dental implants provide an excellent treatment option for patients with missing teeth via the replacement of tooth roots with a fixed permanent artificial root to match the natural ones and support prosthetic dental crowns. Despite the high success rate, implant failures are still common. In this context, long-term implant failure is generally a result of a severe bone marginal loss and bone resorption related to secondary infection . AdditioDue to the limitation of the included studies in this systematic review, it is important to note that none of these studies had indicated a sample randomization process, which might reflect a lower score on the bias scale of funnel graph analysis. Some of the selected studies presented initial sample randomization; however, a smaller number of the patients from those samples were chosen for their study and control groups ,7,17,33.IL genes. These results need to be analyzed with caution, since only two studies were considered in the sample: IL1B  [IL1B (C511T) [IL1B on failure of dental implants.Bone peri-implant marginal loss can be aggravated by chronic or systemic diseases such as cardiovascular diseases, osteoporosis, diabetes, hepatitis, severe periodontal disease, chemo or radiotherapy, human immunodeficiency virus (HIV) positive infection, pregnancy or lactation, or even large bone reconstructions that compromise bone regeneration and integration. The included studies used in this systematic review selected patients who were healthy and with no systemic diseases ,17,18,20rs16944) , IL1B   analyzedorphisms . Anotherorphisms ,20. The 05  analyorphisms  includedorphisms . IL-4 gene was associated with susceptibility of implant loss, when a SNP was included in the analysis of haplotype IL-4, a statistical difference was not identified [Two clinical studies justified the lack of significant evidence due to sample size ,20. In fentified . It emphMultiple published studies highlighted the importance of considering the levels of marginal bone loss around dental implants for future clinical trials . FurtherIL1B, IL10 and TNFα in study groups compared to control groups. Hence, there is no strong evidence that genetic factors can lead to the failure of oral rehabilitation with dental implants. It is important that additional randomized controlled studies with large sample size be conducted in order to conclusively determine whether there is a possible effect of genetic risk factors on implant failure and on marginal bone loss. These findings will help to identify those individuals with higher risk for dental implant loss facilitating the preparation of prevention strategies and individualized therapies in order to increase survival rates of oral rehabilitation with dental implants [In conclusion, the meta-analysis of the combined genetic studies did not show increased risk or protection in dental implant failure due to DNA variations in implants ,4,6.


Although chorioamnionitis (CA) is a well-known risk factor for white matter disease of prematurity, the association with intraventricular hemorrhage (IVH) is controversial and has not been yet systematically reviewed. We performed a systematic review and meta-analysis of studies exploring the association between CA and IVH. A comprehensive literature search was conducted using PubMed/MEDLINE and EMBASE, from their inception to 1 July 2017. Studies were included if they examined preterm infants and reported primary data that could be used to measure the association between exposure to CA and the presence of IVH. A random-effects model was used to calculate odds ratios (OR) and 95% confidence intervals (CI). We found 1,284 potentially relevant studies, of which 85 met the inclusion criteria . Meta-analysis showed that CA exposure was significantly associated with all grades IVH , with grades 1–2 IVH , and with grades 3–4 IVH . Both clinical and histological CA were associated with an increased risk for developing IVH in very preterm infants. In contrast, the presence of funisitis did not increase IVH risk when compared to CA in the absence of funisitis . Further meta-analyses confirmed earlier findings that CA-exposed infants have significantly lower gestational age  and lower birth weight  than the infants not exposed to CA. However, meta-regression and subgroup analysis could not demonstrate an association between the lower GA and BW and the risk of IVH in the CA-exposed infants. In conclusion, our data show that CA is a risk factor for IVH, but also a risk factor for greater prematurity and more clinical instability. In contrast to other complications of prematurity, such as patent ductus arteriosus, retinopathy of prematurity, or bronchopulmonary dysplasia, the effect of CA on IVH appears to be independent of CA as causative factor for very preterm birth. Germinal matrix hemorrhage–intraventricular hemorrhage (GMH-IVH) is one of the most common complications of prematurity Ballabh, . IVH typAs extensively reviewed by Inder et al.  the pathSeveral studies suggest that IVH is unequally distributed among the different leading causes of very preterm delivery DiSalvo, . An estiThe methodology followed the same structure as earlier meta-analyses on CA and ROP  or low BW , as well as studies which used stricter inclusion criteria. Studies were included if they reported primary data on the association between CA-exposure and IVH. We included studies which reported the rate of IVH in infants with and without CA, and studies which reported the rate of CA in infants with and without IVH. The results of the total search were screened independently by two reviewers , in several rounds: first by title only, second by title and abstract, and thirdly by consulting the full text. The reviewers resolved discrepancies in inclusion through discussion and by consulting a third reviewer (P.D).Using a predetermined worksheet, two researchers  extracted data from the studies included. Another two investigators  checked the extracted data for accuracy and completeness. We resolved discrepancies by discussion and through checking the primary report. The following data were extracted from each study: citation information, location of study, primary objective, criteria for inclusion/exclusion of infants, definitions used for CA and for IVH, infant baseline characteristics in the total group and the CA-exposed and CA-unexposed groups, and reported results on the outcomes of interest .We used the Newcastle-Ottawa Scale (NOS) for cohort or case-control studies to assess the methodological quality of included studies. Three aspects of a study are evaluated by the NOS: selection, comparability and exposure/outcome, and these are scored individually and tallied up to a total of 9 points. Two researchers (E.V.-M. and E.V.) independently used the NOS to evaluate the quality of each study, and discrepancies were discussed and resolved by consensus.comprehensive meta-analysis V 3.0 software . We calculated the odds ratio (OR) and 95% confidence intervals (CI) for dichotomous outcomes from the data extracted from the studies. We calculated the mean difference (MD) and 95% CI for continuous outcomes. We used the method of Wan et al.  to evaluate covariates which may affect the effect size , for not defining IVH clearly (k = 9), for only adjusting the risk of IVH for one confounding factor (k = 7), for not defining CA clearly (k = 6), and for losing a substantial portion of infants to follow-up (k = 4).A summary of the quality assessment of each study using the NOS is shown in Supplementary Table Meta-analysis showed that CA exposure was significantly associated with all grades IVH Figure , with grTo confirm findings from earlier reports  in GA from the control group, CA was a risk factor for all grades IVH but not for grades 3–4 IVH  in the development of IVH, we carried out further meta-analyses. Thirteen studies reported on IVH (Ohyama et al., Thirteen studies adjusted the association between CA and the risk of IVH for confounding factors. As shown in Supplementary Tables The current systematic review and meta-analysis demonstrates that both clinical and histological CA are associated with an increased risk for developing IVH in very preterm infants. In contrast, the presence of funisitis did not increase IVH risk when compared to CA in the absence of funisitis. We found through additional meta-analyses that CA-exposed infants had significantly lower GA and BW than infants not exposed to CA. However, meta-regression and subgroup analysis could not demonstrate an association between the lower GA and BW and the risk of IVH in the CA-exposed infants. This suggests that the effects of CA on IVH risk might be at least partially independent on the role of CA as an etiological factor for very preterm birth.The association between CA and increased risk of IVH is biologically and clinically plausible. IVH generally occurs within the three first days of life and affects the infants with higher hemodynamic and respiratory instability, frequently associated with extreme prematurity and/or severe perinatal infections (Mohamed and Aly, The degree of prematurity is the most important predisposing factor for the occurrence of IVH Ballabh, , as wellBesides GA, several other factors potentially confound the association between CA and IVH. A number of studies provided data adjusted for confounding factors, but confounders accounted for in each model differed across studies. We performed separate analyses aggregating adjusted association measures. This reduced or made non-significant the association between CA and IVH (Supplementary Tables Adjustment for potential confounders, particularly for GA and/or BW, is a common strategy used in observational studies analyzing predictors of outcomes of prematurity (Gagliardi et al., Our data suggest that CA-exposed infants are not only younger but also more clinically unstable than the non-exposed infants. This is reflected in the higher mortality and the higher rate of sepsis and PDA in CA-exposed infants Table . Of noteThe biological plausibility of the association between CA and IVH is supported by the direct and indirect effects of inflammatory mediators. Hemodynamic disturbances in preterm infants with CA have been correlated with elevated cord blood concentrations of proinflammatory cytokines such as IL-6, IL-1beta and TNF-alpha (Yanowitz et al., We have discussed the role of funisitis in earlier meta-analyses on CA and ROP (Villamor-Martinez et al., Our meta-analysis has several limitations that should be considered. Firstly, there was substantial heterogeneity in how CA was defined in studies. The definitions of clinical CA in particular varied substantially, and recent recommendations propose restricting the term CA to pathologic diagnosis (Higgins et al., per se, the 1–4 grading system remains pervasive in the literature and clinical setting despite debate regarding appropriate nomenclature (Leviton et al., A significant limitation in any meta-analysis on IVH is the potential for heterogeneity in defining the condition. The grading system most commonly used for neonatal IVH was first reported by Papile et al. and later modified by Volpe and is based on the presence and amount of blood in the germinal matrix, lateral ventricles, and periventricular white matter Volpe, . Grade 1IVH is a multifactorial complication that is more common in more preterm and more clinically unstable infants. We established for the first time through meta-analysis that CA is a risk factor for IVH. We also confirmed earlier findings that CA is a risk factor for being born more preterm and presenting more clinical instability. However, in contrast to other complications of prematurity, such as PDA, ROP, or BPD (Hartling et al., https://dataverse.harvard.edu/dataset.xhtml?persistentId=doi%3A10.7910%2FDVN%2FJ9RHUF.The datasets generated and analyzed for this study can be found in the Harvard Dataverse (Harvard Dataverse, https://www.biorxiv.org/content/early/2018/05/30/334375.A preprint version of this manuscript is made available to the scientific community on the preprint server bioRxiv (Villamor-Martinez et al., EV-M carried out data collection, carried out statistical analyses, assessed methodological quality, contributed to interpretation of results, drafted the initial manuscript, and reviewed and revised the manuscript. MF contributed to the design of the study, the statistical analysis and interpretation of results and reviewed and revised the manuscript. OM selected studies for inclusion, carried out data collection and carried out statistical analyses. SP contributed to interpretation of results and reviewed and revised the manuscript. GC contributed to interpretation of results and reviewed and revised the manuscript. PD carried out and supervised data collection and contributed to interpretation of results. FM contributed to interpretation of results and reviewed and revised the manuscript. EV conceptualized and designed the study, carried out the search and selected studies for inclusion, supervised data collection, contributed to statistical analyses and interpretation of results, and reviewed and revised the manuscript. All authors approved the final manuscript as submitted.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


The tumor immune infiltrate, as recently evaluated with the immunoscore methodology, has been reported to be related to colorectal cancer (CRC) progression. Nevertheless, results varied from different studies. A meta‐analysis was conducted to solve this problem. We collected data from included studies to evaluate the prognostic role of immunoscore in CRC patients on overall survival (OS) and disease‐free survival (DFS). MEDLINE, EMBASE, and Cochrane libraries were searched through 30 June 2018. Hazard ratio (HR) with 95% confidence intervals (95% CI) was pooled using a random‐effects model for OS and a fixed‐effects model for DFS. Finally, eight studies (involving 4689 CRC cases) were identified as eligible publications. The results of the meta‐analysis showed that low immunoscore was significantly correlated with poor OS  and DFS . The findings from most subgroup analyses were consistent with those from the overall analysis. The immunoscore could be a useful prognostic marker in patients with CRC. It is necessary to evaluate immunological markers in international multicenter studies. The system is of much importance but gives limited information for the prognostic benefit of the selected therapy.+) with cytotoxic (CD8+) and memory (CD45RO+) phenotype within the core of the tumor (CT) and the invasive margin (IM) was a highly significant parameter to predict recurrence and survival. To promote the use of this immune investigation as a routine testing for cancer classification, Galon+ and CD8+ lymphocytes in the CT and IM regions of tumors. Patients were stratified from I0 to I4 according to the “immunoscore” (“I”), based on the total number of observed high densities (CD3+ cells and CD8+ cells in the tumor regions).+ and CD8+ in the core of the tumor (CT) and the IM regions of the tumor (4‐Hi); I4 refers to tumors with high densities of CD3+ and CD8+ cells in both tumor regions. In addition, some studies have shown that the immunoscore method was better than the current TNM staging system, especially for colon cancers.Anitei MGTherefore, we conducted the meta‐analysis to assess the effects of immunoscore on overall survival (OS) and disease‐free survival (DFS) in CRC patients.22.1MEDLINE, EMBASE, and Cochrane libraries were searched to find relevant publications up to 30 June 2018, using the search terms “immunoscore OR immune score,” “colorectal cancer OR colorectal neoplasms OR colorectal carcinoma,” “prognostic OR prognosis OR survival,” and combinations thereof. This meta‐analysis was performed under the guidelines of PRISMA.2.2All papers were reviewed by two authors (TXL and QH) independently. Uncertainties and discrepancies were resolved by consensus after discussing with a senior researcher (SGR). The selected studies had to meet the criteria as follows: (a) English articles; (b) pathologically diagnosed CRC patients; (c) OS or DFS of CRC as the research focus; and (d) reporting hazard ratio (HR) estimates with their corresponding 95% CI . When studies were reported in duplication, only the study with the largest sample size and detailed information or the study that met the above criteria was included. Abstracts were excluded. Review articles and editorials were included if they contained original data.2.3Data extraction from each paper was performed by two authors (TXL and QH), and discrepancies were resolved by consensus. The following data were collected from include papers: first author, publication year, country, enrollment time, follow‐up time, characteristics of the study population , and HR estimates with corresponding 95% CIs for OS or DFS HR from multivariate Cox proportional hazard analysis was preferred in the analysis.2.4Q testP value and quantified using the I2 statistic. If I2 < 50%, the fixed‐effects model (Mantel‐Haenszel method)The meta‐analysis was performed, and HR with corresponding 95% CI was calculated in CRC patients. The Cochran’s P < 0.05 was considered statistically significant.Stata/MP version 12 for Windows  was used for all statistical tests. A two‐tailed 33.1Figure 3.2Table 3.33.3.1Pheterogeneity = 0.002, I2 = 67.1%)   for Europeans, but not significantly correlated with Asians . When cancer cases stratified by clinical stage, the low immunoscore was much more significantly correlated with poor OS for IV stage patients  than for I‐III stage patients .The associations of immunoscore and DFS in CRC patients did not differ by number of patients and HR. When cancer cases stratified by max follow‐up time, the low immunoscore was much more significantly correlated with poor DFS for more than 10 years . When cancer cases stratified by clinical stage, the low immunoscore was significantly correlated with poor DFS for I‐III stage patients , which had a similar effect for all‐stage patients.In short, the estimated heterogeneity for studies included decreased to some degree but did not obliterate.3.5To assess the impact of each single study on the pooled HR, we removed individual studies in turn from our meta‐analysis. Figure 3.6Begg’s test (0.095), Egger’s test (0.065), and the near‐symmetric funnel plot Figure  suggeste4+ and CD8+ T‐lymphocyte density, significantly correlated with OS and DFS, suggesting that immunoscore may be new prognostic and predictive marker for CRC.Both clinical characteristics of patients and biological features of the tumor are increasingly used and may be very helpful in the management of cancer disease. This systematic review and meta‐analysis is the first one investigating the effects of immunoscore on OS in CRCs. Our study demonstrated that the immunoscore, an immunohistochemistry‐based assessment of CD3+ and CD8+ lymphocytes is associated with less aggressive clinicopathological features and a better prognosis. However, these conclusions have no significant impact on both cancer classification and clinical decision making. Since then, increasing studies have assessed the association between immunoscore and CRC prognosis. However, the potential association between immunoscore and CRC prognosis remains controversial. So, we were interested in the influence of immunoscore on therapy outcomes of OS and DFS in CRC. The study showed that low immunoscore CRCs were related to a 74% reduction in OS and 82% reduction in DFS, compared with high immunoscore CRCs. These findings supported that TILs play an useful role in boosting anti‐tumor immunity against CRC.Since the beginning of the 20th century, the immune infiltration of cancer has been suspected as a positive factor in the prognosis of patients.Subgroup analysis stratified by geographic region showed that the low immunoscore was significantly correlated with poor OS for Europeans, but not significantly correlated with Asians. The results suggest that the effect of immunoscore for CRC prognosis may have a ethnic difference. Considering that the results of the Asian population study are only from two studies, future studies are encouraged to investigate the difference in survival between different races in CRC.Clinical stage at diagnosis is the most important prognostic factor for CRC.We noticed that one study in our meta‐analysis included a half number of all patientsThe strengths of the present study include the following: (a) The present analysis is the first to investigate the effects of immunoscore on OS in CRCs; (b) it included a large sample size (4689 CRC cases); (c) to minimize the effect of potential confounders, a strict inclusion criterion, advanced meta‐analysis of HR for survival, and fully outcomes of interest (OS and DFS) were adopted; (d) subgroup analyses stratified by the geographic region, number of patients, max follow‐up time, clinical stage, pathological type, and HR Thus, the effect of potential confounders was minimized; and (e) the results of sensitivity analysis and no publication bias were observed in the analyses, indicating the results are robust.However, our meta‐analysis has some limitations, including (a) Significant heterogeneity was found among studies when pooling the HRs for OS. To solve this problem, the meta‐analysis of random‐effects model was used to pool data whenever significant heterogeneity was noted. Appropriate well‐motivated inclusion criteria were used to maximize homogeneity, and sensitivity analysis was performed to investigate potential sources of heterogeneity; (b) in this meta‐analysis, the studies selected were limited to English papers and three universal used databases, which may lead to language and selective bias; and (c) subgroup analysis was limited for less studies included, especially for PFS.In summary, the findings of our meta‐analysis highlight the performance of the immunoscore in predicting the clinical behavior of patients. It is time to begin evaluating immunological markers in international multicenter studies.No potential conflicts of interest were disclosed.


Steroid treatment has become recognized as an important risk factor for avascular osteonecrosis of the femoral head. However, not all patients who receive long-term, high-dose steroids develop osteonecrosis, indicating that there are individual differences in occurrence.We explored the relationship between polymorphisms and steroid-induced osteonecrosis of the femoral head (SONFH) incidence with variables. We used a multilevel mixed-effects logistic regression model, which is an expansion of logistic regression, for each type of steroid, primary disease, drug dose, applied duration, and single-nucleotide polymorphism (SNP). We also conducted a dose-response meta-analysis to analyze the cumulative dosage and SONFH risk in mutation carriers. There were significant correlations between the ABCB1 rs1045642 mutant and SONFH in the prednisone-use and methylprednisolone/prednisone-use populations. The ABCB1 rs2032582 mutant homozygote had a protective effect in the methylprednisolone/prednisolone renal transplant population. For ApoB rs693, mutation increased the incidence of SONFH in prednisone-use and methylprednisolone/prednisolone-use populations and renal transplant patients. For ApoB rs1042031, mutation increased the risk of SONFH in the prednisone-use population. The PAI-1 rs1799768 mutation had a protective effect on the SONFH risk prednisone-use and renal transplant populations. ABCB1 rs1045642 mutations have a protective effect against SONFH, and ApoB rs693 and rs1042031 increase the SONFH risk. Cumulative dosage and treatment duration had little effect on the results. In addition, there was a dose-effect correlation in ABCB1 rs1045642 and rs2032582 mutation carriers. Steroid-induced osteonecrosis of the femoral head (SONFH), which leads to collapse of the femoral head and articular dysfunction, has an incidence of 9–40% amongst patients receiving steroid treatment . The exaTo date, several meta-analyses have been published and shown that PAI-1 4G/5G rs1799768)  and CYP3  and CYPWe used the meta-analysis of observational studies in epidemiology guidelines in the present study .Two authors independently performed a literature search of the PubMed, Embase, Cochrane Library, and Chinese public databases, including the China National Knowledge Infrastructure, the China Biology Medicine Database, the China Science Periodical Database (Wanfang Database), and the VIP Journal Integration Platform. The search included studies published through 29 July 2018. The following terms were used in the search strategy: hormone, glucocorticoid, steroid, corticosteroid, osteonecrosis, femoral, femur, femoris, whirlbone, polymorphism, SNP, genetic, mutation, genotype, allele, allelic, and variation. The search strategy is shown in Supplementary Table S1. Divergence in the search results was resolved by discussion.The studies were included in our meta-analysis if they met the following criteria: (1) case-control or cohort studies comparing a population that suffered SONFH with a population that did not suffer after steroid treatment, (2) studies assessing the associations between genetic polymorphisms and SONFH, and (3) studies reporting the frequencies of specific alleles or the effect sizes of individual genotypes between cases and controls. Studies were excluded from the analysis for the following reasons: (1) noncase-control or noncohort studies, (2) the case group included SONFH patients with other etiologies or SONFH patients who were not reported separately, (3) the control group included an ONFH population without steroid application or a healthy population, (4) non-SNP-related studies, (5) studies about family heredity, and (6) studies that did not report data pertaining to allelic frequencies or effect size. In addition, conference reports, editor comments, reviews, case reports, and academic dissertations were excluded from the analysis.2 and the average body weight as 60 kg. For closed-interval categories, the value assigned to each dosage or duration, such as the average or median, was the midpoint. For open-interval categories, the value was estimated as the limit value multiplied by 1.2 [Two authors independently extracted the following data from each eligible study: first author’s name, publication year, research location, sample size, average subject age, primary disease, steroid type, cumulative steroid dose, treatment duration, and genes of interest. If the average or median value of the cumulative steroid dosage in each group was not reported, it was estimated. The cumulative dose was generally calculated by multiplying the daily average dose and the treatment duration. For studies with a fixed treatment strategy, the cumulative dosage was calculated by multiplying the dosage of each treatment cycle by the total number of cycles. In the present study, we defined the average body surface area as 1.6 md by 1.2 . Becaused by 1.2 . The Newd by 1.2 .2 statistic was used to estimate the degree of heterogeneity amongst the studies. Cochran’s Q statistics were also taken as measures of between-study heterogeneity. Generally, a significance level of P<0.1 suggests the existence of significant heterogeneity to expand a test’s sensitivity [2 ≥ 50% , the random-effect model was used; otherwise, the fixed-effect model was used. Then, we attempted to use a multilevel mixed-effects logistic regression model, which is an expansion of logistic regression, for each type of steroid, primary disease, drug dose, applied duration, and SNP. The factors of different genotypes and other variables were included as fixed effects, and different studies were considered random effects.Association analysis was performed using five genetic models: allelic (W vs M), dominant (WW+WM vs MM), recessive (WW vs WM+MM), heterozygous (WM vs MM), and homozygous (WW vs MM) models . W is thsitivity . If the Finally, we conducted a dose-response meta-analysis across the cumulative steroid dosage . To deriOur research returned 287 English articles and 305 Chinese articles after removing duplicates. After screening the titles and abstracts, 509 of these articles were excluded. The full texts of 83 articles were assessed, amongst which studies were excluded for the following reasons: the control group did not receive steroid therapy (26); reviews (8); non-SNP-related studies (5); basic studies (3); studies without frequencies or effect size results (2); non-steroid-related ONFH studies (2); case reports (2); microRNA-related studies (2); duplicated reports (2); and heredity SONFH study (1). Ultimately, we included 30 studies that assessed a total of 7553 patients in our meta-analysis  Figure , Table 1All studies were case-control designs, except one study, which was a cohort design . The priP=0.046). ApoB rs693  and rs1042031  mutations increased the risk of SONFH. ABCB1 rs2032582, MTHFR rs1801133, and PAI-1 rs1799768 had no significant relationships with SONFH . The results showed no significant changes in the cumulative dosage and treatment duration. According to the analyses of categorical variables of steroid type, in the prednisone-use population, mutant homozygotes had a significant protective effect compared with the wild-type population , and the results were similar in the methylprednisolone/prednisone-use population . The analysis of primary disease showed no significant results. It should be noted that the studies that did not report primary disease type or that included several types of primary disease were not included in the categorical analysis of primary disease . However, this result was based on a single study and had a large standard error (SE = 0.66); thus, this result needs to be confirmed   . Due to P=0.002) (P<0.001), but not in the methylprednisolone/prednisolone-use renal transplant population .For ApoB rs1042031, mutation increased the risk of SONFH  . The resP=0.143; TT vs CC: OR: 1.19; 95% CI: 0.48–2.97; P=0.705; CT/TT vs CC: OR: 0.94; 95% CI: 0.61–1.45; P=0.776). The results did not change after considering the continuous and categorical variables  .For PAI-1 rs1799768, 5G mutation had a protective effect against SONFH  . These rP=0.008; multiple dose: P=0.0064)  and heterozygous mutation patients  . In wild-type populations, there was no correlation between the dose and the risk of SONFH . Although there was no significant correlation between the ABCB1 rs2032582 mutation and the risk of SONFH, the cumulative dose in the mutant population was correlated with the occurrence of SONFH. As the dose increased, the risk of SONFH decreased  , but not in other mutation patients  . In wild-type populations, there was no correlation between the dose and the risk of SONFH . These findings may suggest that mutation carriers still have a risk of SONFH when using steroids, similar to the wild-type population. If SONFH does not occur after small-dose steroid application, increasing the steroid dose in mutation carriers will not increase but rather decrease the risk of SONFH.Furthermore, we sought to analyze the cumulative dosage and SONFH risk in mutation carriers by a dose-response meta-analysis. SONFN/ONFH incidence in mutation carriers cannot be obtained when steroids are not used. Therefore, the minimum dose must be used as a reference, and whether increasing the dose affects SONFH incidence must be assessed. Two dose assessment methods were used: an absolute dose and multiple doses. For ABCB1 rs1045642, there was a correlation between the cumulative dose and the occurrence of SONFH in the mutation carriers. As the dose increased, the risk of SONFH decreased (absolute dose: =0.0064) A,B. This=0.0258) C,D. ThisP=0.0064; multiple dose: P=0.0064), a correlation was also found in the wild-type population . For ApoB rs1042031, there was no correlation between the cumulative dose and SONFH risk in the mutation carriers  or in the wild-type population . ApoB rs693 and rs1042031 have too few homozygous mutation patients to analyze separately. We did not perform a dose-response analysis for MTHFR rs1801133 and PAI-1 rs1799768 due to the small number of included studies.For ApoB rs693, although SONFH was correlated with the dose in the mutation carriers  and 2677T/A (rs2032582) alleles of ABCB1 were identified, which may reduce the risk of GC-induced ONFH 42 and 26. The resIn the present study, there were significant correlations between the ABCB1 rs1045642 mutant and SONFH in the prednisone-use and methylprednisolone/prednisone-use populations. The ABCB1 rs2032582 mutant homozygote had a protective effect in the methylprednisolone/prednisolone renal transplant population. For ApoB rs693, mutation increased the incidence of SONFH in the prednisone-use and methylprednisolone/prednisolone-use populations and renal transplant patients. For ApoB rs1042031, mutation increased the risk of SONFH in the prednisone-use population. The PAI-1 rs1799768 mutation had a protective effect in the SONFH risk prednisone-use and renal transplant population.There were different outcomes for the different types of steroid-use populations, possibly because the results were based on even fewer studies. These differences may also indicate that different types of steroids have subtle differences in affecting the risk of SONFH. It is generally believed that prednisone must be transformed to an active form through the liver, while prednisolone does not require this change, which may be the reason for the different results of the two drugs in ABCB1-related mutation states. In addition, more evidence is needed to determine whether the different results for the types of steroids in ApoB and PAI-1 gene polymorphisms were due to different effects on adipocytes.Currently, a registered study is analyzing the pharmacokinetics of different steroids in children, but the results have not yet been reported (NCT02252237). An earlier study also reported that the cumulative dosage of methylprednisolone, but not prednisone, is correlated with avascular necrosis incidence . TherefoAlthough there are reports about the effect of primary disease on SONFH risk, most have focussed on the influence of steroids used in disease treatment. It was also reported that in renal transplant patients, cyclosporine versus tacrolimus as well as gender factors may be independent risk factors for avascular necrosis . A meta-It is generally believed that the occurrence of SONFH is due to long-term and/or high-dosage steroid application. A dose-response meta-analysis showed that the incidence of osteonecrosis is correlated with both the cumulative dose and treatment duration . HoweverThe present study still has several limitations. First, our study was performed at the study level instead of the individual level. Second, in stratified analyses, with the increase in variable factors, the results become more dependent on the results of single studies, which will reduce the stability of the results. Third, in dose-response analyses, SONFN/ONFH incidence in mutation carriers without steroid application cannot be calculated; therefore, the minimum dose must be used as a reference. Fourth, our study only analyzed SNPs that were examined in more than three studies, but with emerging research and evidence, more SNPs may be studied in the future. Therefore, the present study was limited by the research available at the time.


Numerous studies have explored the correlation of periodontal disease (PD) with risk of hematopoietic and lymphatic cancers, but the findings were inconsistent. Therefore, we did a meta-analysis to ascertain the correlation of PD with risk of incident hematopoietic and lymphatic cancers.  The authors searched relevant studies in databases . The summary relative risk (RR) along with 95% confidence interval (CI) was calculated by use of random or fixed effects models. P = 0). Stratified analysis showed the association of PD with hematopoietic and lymphatic cancers remained significant in the never smokers , and in the American population , respectively. Six studies were included in qualitative synthesis. The pooled analysis revealed that PD was significantly associated with an increased risk of hematopoietic and lymphatic cancers (RR = 1.17; 95% CI = 1.07–1.27;  Never smokers population and the American population with PD have a higher risk of developing hematopoietic and lymphatic cancers. PD might be considered as a risk factor for hematopoietic and lymphatic cancers.  Key words:Periodontal disease, hematopoietic and lymphatic cancer, meta-analysis, systematic review. PD is a chronic inflammatory condition that affects the tissues surrounding the teeth, including alveolar bone, cementum, periodontal ligament and gingival. If periodontitis remains untreated, it may lead to annihilation of periodontium, tooth looseness and ultimately tooth loss. PD has been reported to increase the risk of chronic diseases, such as diabetes, cardiovascular disease, rheumatoid arthritis, and pulmonary disease -5. FurthHematopoietic and lymphatic malignancies are cancers that affect blood, lymphatic system, and bone marrow. There are a number of risk factors associated with lymphoma , such as age, gender, Epstein-Barr infection, and immune dysregulation (https://www.cancer.gov/types/lymphoma). The most common risk factors associated with hematopoietic cancers  include older age, smoking, prior chemotherapy, and exposure to radiation (https://www.cancer.gov/types/leukemia).Recently, numerous epidemiological studies exploring the risk of hematopoietic and lymphatic cancers in patients with periodontitis have been published -19. Howe- Literature and search strategyWe followed the instructions of the PRISMA to conduct this meta-analysis . The focPubMed, Web of Science, and MEDLINE were systematically searched to identify potentially eligible studies that assessed the correlation of PDs with hematopoietic and lymphatic cancers. We used the search terms:  and  and . The search was completed on 31 Oct 2018.- Selection criteria, data extraction and quality assessmentPublished studies that estimated the relative risk of hematopoietic and lymphatic cancers associated with any measure of PD  were eligible. Comments, review articles and meta-analysis were excluded.The following data from each included study were independently extracted by two investigators (YGW and XJS). The relevant data included first author of the study, year of publication, country of origin, study design, sample sizes in each group, age, gender, cancer type, duration of follow-up, exposure, relative risk (RR), hazards ratios (HR), or odds ratio (OR) estimates with 95%CI for outcomes.The quality of included studies was assessed according to the Newcastle-Ottawa Scale (NOS) . This sc- Statistical analysisRelative risk (RR) with 95% confidence interval (CI) was used as the common measure of association across studies. RR was calculated by the formula RR = exposure/non-exposure, in which exposure and non-exposure is the incidence of an outcome of interest in the exposed group and non-exposed group, respectively. HR was considered equivalent to RR when pooled in qualitative synthesis. When the incidence of outcome is more than 10% in a cohort study, the OR was transformed into RR with this formula: RR = OR/[(1 – P0) + (P0 × OR)], in which P0 is the incidence of the outcome of interest in the non-exposed group . The staP < 0.10 was regarded as statistical significance) and was quantified with I2 statistic. I2 values higher than 50% indicated moderate heterogeneity. In the presence of heterogeneity (P < 0.10), we combined the effect estimates across studies by use of a random-effects model . No evidence of heterogeneity was observed  but not among past  or current smokers . A gender-specific analysis showed that the association between PD and risk of hematopoietic and lymphatic cancers was not statistically significant in male and female population , and female population , respectively. Four studies were performed in America, one in Asia, and one in Europe. For subgroup analysis by geographic region, the significant association between PD and risk of hematopoietic and lymphatic cancers was observed among the American population . Stratification analysis by ethnicity demonstrated no significant association between PD and risk of hematopoietic and lymphatic cancers in the white population or non-white population. Moreover, pooled estimates found no evidence of increased risk of hematopoietic and lymphatic cancers among individuals with mild/moderate PD  and severe/edentulism PD , respectively.- Sensitivity analysesIn sensitivity analysis using the leave-one-out approach, the results showed that no individual study significantly altered the pooled RR for PD on hematopoietic and lymphatic cancers risk, with pooled RR varying from 1.12  to 1.20 , demonstrating the meta-analysis results were robust .- Bias assessmentP = 0.224).To examine the extent of publication bias for included studies, funnel plot and Egger regression analysis were conducted. The funnel plot on included studies showed a little asymmetry Fig. . The EggThe results on the association between PD and the risk of hematopoietic and lymphatic cancer are less consistent. No associations were detected in four of the six included studies ,15,17,18et al. found that anti-inflammation therapies of PD did not reduce the risk of hematopoietic and lymphatic cancers . Four studies were prospective cohort studies -18, whicIn summary, our pooled analysis indicates that periodontitis is significantly associated with an increased risk of hematopoietic and lymphatic cancers, especially lymphatic cancers. Moreover, the associations are particularly strong in the population of never smokers and the American population. However, additional prospective studies with a larger inclusion population, improved measurement and classification of PD, longer follow-up duration, and well-designed clinical models adjusted for the major and common confounders  are warranted to confirm present findings. Additional research is expected to assess whether PD prevention and treatment can decrease the incidence of hematopoietic and lymphatic cancers. Findings of this meta-analysis highlight the importance of maintaining good oral health, which may led to a reduced risk of developing hematopoietic and lymphatic cancers.


Furthermore, we observed similar negative results in the association test of subgroup analysis . Heterogeneity, Begg's, and Egger's tests excluded the presence of high heterogeneity and publication bias. Statistically stable results were observed in the sensitivity analyses. Based on integrated analysis of the current evidence, Prothrombin gene G20210A polymorphism is likely unrelated to the risk of RVO.The aim of this study was to perform an updated meta-analysis to quantitatively investigate the association between G20210A polymorphism of Prothrombin gene and the risk of retinal vein occlusion (RVO), based on the available publications with inconsistent results. We utilized the Stata software to perform the heterogeneity test, association test, Begg's and Egger's tests, and sensitivity analysis. We searched three on-line databases  and obtained a total of 422 articles. Based on our selection criteria, 24 case-control studies were finally enrolled in this overall meta-analysis; a subgroup analysis by the factors ethnicity, control source, and RVO type was done. Through the association test of overall meta-analysis, we did not observe a significant difference between RVO cases and controls under the A  Retinal vein occlusion (RVO) is a common retinal vascular disease, and often contributes to the occurrence of visual decline or loss, especially for middle-aged or elderly individuals . The maiFactor V G1691A (Factor V Leiden or R506Q) and G20210A polymorphism (rs1799963) within Prothrombin (Factor II) gene are the most common inherited thrombophilic mutations . PreviouTo the best of our knowledge, only two meta-analyses on the genetic role of Prothrombin G20210A in the susceptibility to RVO were reported in 2005 . In the Three authors  performed the on-line database search (updated to November 2018) to obtain the related published articles. Three on-line databases, including PubMed, Excerpta Medica Database (Embase), and Web of Science (WOS), were electronically searched. No restrictions of publication period or language were utilized. Detailed search terms are shown in Supplementary Table S1.Three authors  selected the eligible case-control studies. Based on the principles of PICOS , the specific inclusion criteria were utilized: (P) cases of RVO; (I) Prothrombin G20210A polymorphism; (C) healthy individuals or negative controls; the genotype frequency distribution should follow Hardy-Weinberg equilibrium (HWE); (O) “GG”, “GA”, and “AA” genotype frequency data of G20210A polymorphism in both cases and controls; (S) case-control studies. Articles were removed according to our specific exclusion criteria, which were duplicate studies, other disease or gene, cell or animal data, review or meta-analysis, meeting abstract, case report or trial, and lack of confirmed genotype data. When encountering disagreements, a discussion with another author (S. Zhao) took place for a final consensus.Three authors  extracted the data from the eligible case-control studies. A form was utilized to summarize the characteristics, including the first author, publication year, country, ethnicity, genotype frequency, control source, genotyping assay, and sample size. When genotype frequency data was missing or unavailable, we tried to contact the author through an e-mail.2 value less than 50% indicated the existence of heterogeneity between studies, and a fixed-effect model was used in the association test (Mantel-Haenszel statistics). Four inheritance models, including A vs G , A vs G (carrier), GA vs GG (heterozygote), and GA+AA vs GG (dominant), were utilized. The odds ratios (OR), 95% confidence intervals (CI), and P values of association tests were determined.Overall meta-analysis and subgroup analyses by three factors, including ethnicity, control source, and RVO type, were performed using Stata software . In the heterogeneity test, a P value of Cochran's Q statistic larger than 0.1 or IBegg's and Egger's tests were performed to assess the potential publication bias. P values less than 0.05 indicated the existence of potential publication bias. In addition, sensitivity analysis was performed to evaluate the stability of statistical results.After the database search, we identified a total of 422 articles  and removed the 100 duplicate articles. Then, we excluded another 292 improper articles, according to our exclusion criteria . Of the remaining 30 articles, six articles were excluded because no confirmed genotype data in both case and control groups were obtained. As a result, a total of 24 eligible case-control studies 0, meetin were enr2 =0.0%, vs G  , A vs G (carrier) , GA vs GG , and GA+AA vs GG . vs G  model.There were a total of 24 case-control studies in the overall meta-analysis. The absence of heterogeneity  model. These findings suggested that G20210A polymorphism within Prothrombin gene had no role influencing the risk of central or branch retinal vein occlusion in the Caucasian population.Next, we performed the subgroup analyses by the factors of ethnicity (Caucasian), control source , and RVO type (BRVO or CRVO). As shown in vs G  model. Additionally, we observed a relatively stable conclusion through the sensitivity analysis , rather than BRVO (P=0.09) . HoweverIn 2005, Janssen et al. included six case-control studies ,27,29,30vs GG) and recessive (AA vs GG+GA) models. The same data was obtained in the heterozygote and dominant models. Although we utilized a new selection strategy and added some newly published case-control studies, no significant association between Prothrombin G20210A polymorphism and RVO risk was obtained in our updated pooling analysis.In order to enroll the maximum number of eligible case-control studies, we systematically searched the three on-line databases  and applied a strict selection criteria. In comparison with the two prior reports, we ruled out one case-control study without Prothrombin G20210A mutation  and threPAI-1) gene and C677T (rs1801133) polymorphism of 5,10-methylenetetrahydrofolate reductase (MTHFR) gene, in the risk of ROV needs to be evaluated.Our sensitivity analyses data indicated the statistical robustness of pooling results while the heterogeneity Begg's and Egger's tests data supported the absence of high heterogeneity or publication bias. In spite of this, we should consider the existence of limitations within our meta-analysis. First, just like other meta-analyses, the statistical power of our pooling analysis was affected by the small number of enrolled studies. For instance, only one case-control study was enrolled in the Asian subgroup analysis . The negProthrombin G20210A polymorphism was associated with enhanced susceptibility to venous thromboembolism (VTE), especially deep venous thrombosis and pulmonary embolism –36. MorTaken together, our updated meta-analysis did not statistically support the genetic correlation between Prothrombin gene G20210A polymorphism and the risk of central or branch retinal vein occlusion. However, additional case-control studies in different populations are still needed.pdf].Click here to view [


GSTM1 and GSTT1 are involved in detoxification of xenobiotics, products of oxidative stress and in steroid hormones metabolism. We investigated whether GSTM1 and GSTT1 gene deletion was associated with DTC risk and explored interaction with non-genetic risk factors of DTC.The study included 661 DTC cases and 736 controls from two case-control studies conducted in France and New Caledonia. Odds ratios (OR) and their confidence interval (CI) for DTC associated with GST genotypes, alcohol drinking, tobacco smoking, body mass index and hormonal factors were calculated using logistic regression models.GSTT1 and GSTM1-deleted genotype, and 1.26, (95%CI 0.89–1.77) in carriers of both genes (p-interaction = 0.02). The OR for drinking ≥1 glass/week was 0.33 (95%CI 0.15–0.74) in GSTT1-null individuals while it was 1.01 (95%CI 0.67–1.52) in non-null carriers of the gene (p-interaction = 0.01). No interaction between GST genotypes and other non-genetic risk factors was detected.Results are presented for Europeans and Melanesians combined, as no heterogeneity between groups was detected. We found that DTC risk increased with obesity and decrease with alcohol drinking. After stratification by gene deletion status, the OR for obesity was 5.75, (95%CI 2.25–14.7) among individuals with GSTM1 and GSTT1 genotypes may modulate the DTC risk associated with BMI and alcohol consumption. Differentiated thyroid carcinoma (DTC) accounts for about 90% of all thyroid cancers. The incidence of thyroid cancer is characterized by 4–5 times higher rates in women than in men and considerable ethnic and geographic variation ,2. PartiApart from exposure to ionizing radiation during childhood, a well-established risk factor for DTC, increased DTC risk has also been associated with overweight, iodine deficiency, high parity, and late age at menarche. By contrast, several studies have reported that the risk of DTC decreased with alcohol or tobacco consumption –11. ThyrGSTM1 and GSTT1, encode for phase II enzymes, and are involved in the steroid hormone metabolism and in the detoxification of various xenobiotics and products of oxidative stress. The enzyme activity depends on the number of copies of GSTM1 and GSTT1  in the genome [Glutathione S-transferase (GST) genes, such as e genome ,15, withe genome .GSTM1 or GSTT1 genotypes, significant heterogeneity between studies was reported [In a meta-analysis of 12 studies conducted in countries from Europe, Asia and South America that investigated the association between DTC and reported . HeterogSTM1 and GSTT1 genotypes in DTC risk, and their interaction with suspected risk factors of DTC that could be modulated by these genes such as hormonal factors, cigarette smoking and alcohol drinking, and obesity.In the present paper, we used data collected in studies on DTC conducted in populations of European origin in France and in populations of European and Melanesian origins form New Caledonia, to investigate the role of GWe used data from two case-control studies on thyroid cancer conducted in metropolitan France (CATHY study) and in New Caledonia (NC study). All participants provided signed informed consent. The study was approved by the review board of the French institute of health and medical research (INSERM) and authorized by the French data protection authority (CNIL).départements”  covered by a cancer registry. Cases were patients living in these areas aged 25 years and older diagnosed with DTC between 2002 and 2007. Controls were selected at random using the telephone directory and unlisted phone numbers of all private homes in the study areas, and were frequency-matched to the cases by 5-year age group and study area. To prevent possible selection bias arising from differential participation rates across categories of socioeconomic status (SES), the control group was selected to reflect the distribution by SES categories of the general population, as described in details previously [The CATHY study  is a popeviously . From 62The NC study is a country-wide, population-based, case-control study ,7,19. ThIn the two studies, information on ethnicity, personal and familial history of thyroid disease, gynecological and reproductive history, anthropometric factors, diet, alcohol intake, tobacco smoking, and residential and occupational histories was collected during in-person interviews by trained interviewers. BMI was defined as weight (in kilograms) divided by height (in meters) squared. Alcohol drinking was assessed as the lifetime average number of glasses per week using information from the questionnaire for each type of beverage separately . Because the ethanol content is approximately the same for an ordinary glass of any alcoholic beverage, the total number of drinks per week was used as an indicator of total alcohol intake. Pack-years of cigarette smoking were calculated from the total number of years of tobacco smoking and from the number of cigarettes smoked per day for each smoking period.Women were considered to be users of oral contraceptives if they had ever taken pills for at least 6 months. Women were considered postmenopausal if they reported no menstruation for at least 1 year or used menopausal hormone therapy (MHT) , or if they had bilateral ovariectomy . Women with unknown menopausal status, because of hysterectomy before cessation of menstruations or unknown date of last menstruation, were considered postmenopausal if they were 50 years old or more .Association between DTC risk and reproductive factors, BMI, tobacco smoking and alcohol drinking were analyzed in detail previously in the NC , 19 and GSTM1 and GSTT1 using TaqMan gene copy number detection designed by Applied Biosystems. The genotyping was processed by Integragen . Real-time PCR was run on an Applied Biosystems 7900HT Fast system with gene-specific primers for GSTM1 probes (Hs02575461_cn) and GSTT1 probes (Hs00010004_cn), along with primer for the RNase P gene as a reference. Each sample was run in triplicate using 50 ng of genomic DNA. The gene-specific primers were validated in 90 CEPH individuals and genotyping carried out blindly as to case-control status. CopyCaller software V1 (Applied Biosystems) was used to quantify the number of copies in each sample.DNA extracted from saliva samples was used to determine the copy number variation (CNV) of GSTM1 and GSTT1 and were excluded from the analysis. Therefore, a total 1124 Europeans (504 cases and 620 controls) and 270 Melanesians (156 cases and 114 controls) were included in the analyses.Thirty-one subjects (5 cases and 26 controls) in the CATHY study and 14 subjects (8 cases and 6 controls) in the NC study had missing genotypes for both GSTM1 and GSTT1 genotypes using unconditional logistic regression, comparing categories defined by the gene copy number coded as a categorical variable  and adjusting for age , sex, and area of residence. We also compared non-carriers (null genotype) to carriers of at least one copy of the gene (non-null genotype). We first analyzed Europeans and Melanesians separately and tested the heterogeneity of the ORs using the Cochran’s Q test or likelihood ratio test. As no heterogeneity was detected, the combined dataset was used in further analyses .Odds ratios (ORs) were calculated for GSTM1 and GSTT1 genotypes and tobacco smoking, alcohol intake, BMI, and hormonal and reproductive factors using the likelihood ratio test, comparing models with and without the interaction term. Tobacco smoking, alcohol intake, and BMI were analyzed in men and women together, whereas hormonal and reproductive factors were analyzed in women only. To account for the potential confounding effect of hormonal and reproductive factors, we also performed the analysis for tobacco smoking, alcohol drinking, and BMI in women separately. The results were similar to those obtained for men and women combined and are not shown.We tested the interaction between The characteristics of the study sample are presented in Obesity (BMI ≥ 30 kg/m2) and high parity were more prevalent in Melanesians than in Europeans. Conversely, alcohol drinking and use of oral contraceptives were less common in Melanesians . HoweverThe proportion of control subjects with a ‘null’ genotype was much higher in Melanesians than in Europeans  . Odds raGSTM1 and GSTT1 genotypes, using a single dataset combining Europeans and Melanesians as we did not detect heterogeneity between the analyses conducted for each group separately (not shown). The association with obesity was greater among individuals with GSTM1- and/or GSTT1-null genotypes than individuals with non-null genotypes. The OR in obese individuals with null genotype for both genes was 5.75 while it was 1.26 in those with at least one copy of the genes (p-interaction = 0.02). While comparing individuals with null genotype for both genes and BMI≥30 kg/m2 to individuals with non-null genotype and BMI<25 kg/m2, the OR was 2.69 (p = 0.01)  and in those with deletion of both genes . The OR comparing individuals drinking more than one glass per week with non-null genotypes to never drinkers with null genotypes for both genes was 0.36 (p = 0.01) (In  = 0.01) . The inv = 0.01) . The assThe analyses for hormonal and reproductive factors in women stratified by genotypes are shown in GSTM1 and GSTT1-null genotypes than in those with non-null genotypes. A significant interaction between GSTT1-null and age at menarche was also observed. By considering genes and environmental or lifestyle factors, no heterogeneity between groups was observed.This study was conducted in two populations with different genetic background and distinct patterns of exposure to environmental factors. The associations we reported between DTC and lifestyle, hormonal and reproductive factors are consistent with those published previously in the total samples of NC and CATHY studies , 18, 19.GSTM1 and GSTT1 CNVs among Europeans in our study was similar to previous reports [The distribution of  reports ,21. To t reports .GST-null than non-null individuals, particularly when both genes were deleted. This finding was consistent in both ethnic groups. To the best of our knowledge, the only study to investigate the joint effect of BMI and GST genes in DTC was conducted in Korea and found no indication of an interaction [Being overweight has consistently been associated with DTC risk ,24. In teraction . HoweverGSTM1 and GSTT1 contribute to reducing oxidative stress by conjugating and eliminating ROS products [Obesity may increase the risk of DTC through complex mechanisms . The chrproducts , obese sGSTM1/GSTT1-null subjects than non-null subjects, particularly in those with a deletion of both genes. An interaction between alcohol intake and GSTM1 or GSTT1 was reported previously in cancer of the breast, lung, and stomach [The observed inverse association of alcohol drinking with DTC is in line with several studies ,11,29–31 stomach –35. Our The mechanisms underlying the potentially protective effect of alcohol drinking in thyroid carcinogenesis are not known. Some studies have suggested that free radicals generated by alcohol metabolism may have either a direct toxic effect on the thyroid tissues or disturb the hypothalamus–pituitary–thyroid axis ,36,37. HGSTM1 and GSTT1 were not found to modify the association between tobacco smoking and DTC. The interaction between GST genes and smoking was also investigated in a previous study, but no evidence of an interaction was reported [Unlike alcohol, reported .Given the higher incidence of DTC in women than in men, female sex hormones have been suspected to play a major role in thyroid carcinogenesis . We founGSTM1 and age at menarche, use of oral contraceptives, parity, or menopausal status, whereas significant interaction was observed between GSTT1 and age at menarche.Our study is the first to investigate the interaction between GST genotype and reproductive factors in thyroid cancer. No interaction was observed between GSTM1 differs during menstruation [GSTT1 may modulate the association between late age at menarche and DTC risk remains to be elucidated.The association of DTC with age at menarche has been inconsistent in epidemiological studies, and the potential underlying mechanisms involved are unclear. It has been shown that expression of truation  and polytruation . HoweverGSTM1 or GSTT1 genotype have had limited sample sizes of less than 300 cases. Unlike previous studies contrasting DTC risk in null vs. non-null genotypes, CNVs of GSTM1 and GSTT1 were also examined in our study to investigate a potential dose-effect relationship between enzymatic activity and cancer risk. However, such a dose-effect association was not observed in our data. We also examined gene-environment interactions between GSTM1 and GSTT1 genotypes and suspected risk factors of DTC. The analyses of risk factors in these two ethnic groups with diverse prevalence of exposure to risk factors and frequency of GST-null genotypes could help explain the difference in DTC incidence between the study populations. In particular, the highest prevalence of both GST-null genotypes and obesity in Melanesians compared to Europeans could partly explain the higher incidence of DTC in Melanesians.This study used data collected from two case-control studies in Europeans and Melanesians with a population-based design and exhaustive identification of thyroid cancer cases. Our study was based on a relatively high number of subjects. With the exception of one study in Korea that included 1372 cases and 1669 controls , previouOne limitation of our study is the limited statistical power in some strata of the stratified analyses. Potential recall bias inherent to case-control studies may have occurred when assessing exposure to several risk factors. But differential misclassification was unlikely as cases and controls were interviewed in the same conditions by trained interviewers. Also, we cannot exclude that some of the associations found are due to the number of tests used. Replication in independent studies are necessary to confirm our results.GSTM1 and GSTT1 may modify the associations between DTC risk and obesity, alcohol consumption, and possibly hormonal factors. Disparities in the frequency of GSTM1 and GSTT1 deletion and in the prevalence of obesity between populations worldwide, may partly explain differences in thyroid cancer incidence.In conclusion, our results suggest that S1 Table(XLSX)Click here for additional data file.