We found little evidence for sex-specific associations of circulating bilirubin with CRC risk, and further studies are needed to confirm or refute the potential associations.Emerging studies have suggested that bilirubin, particularly unconjugated bilirubin (UCB), has substantial anti-inflammatory and antioxidant properties that protect against oxidative stress-associated diseases such as cancer. Few observational studies have investigated the etiological role of bilirubin in colorectal cancer (CRC) development. In this case-control study, nested in the population-based prospective cohort of the Cooperative Health Research in the Region of Augsburg (KORA) study in south Germany, pre-diagnostic circulating UCB concentrations were measured by high-performance liquid chromatography in 77 CRC cases and their individually matched controls. Multivariable unconditional logistic regression was used to estimate the odds ratios (OR) and 95% confidence intervals (CI) for associations between log-transformed UCB levels (log-UCB), standardized per one-standard-deviation (one-SD) increment, and CRC risk. The models were a priori stratified by sex based on previous evidence. In the fully adjusted models, each one-SD increment in log-UCB was indicative of a positive association with CRC risk  among men, and of an inverse association  among women (P Colorectal cancer (CRC) is one of the leading causes of cancer mortality worldwide. There were over 1.8 million new cases in 2018, and CRC rates are substantially higher in men than women [Oxidative stress, expressed as a constant increased reactive oxygen species (ROS) load, and inflammation are involved in the development of a variety of diseases including cancer . CRC is Emerging studies have suggested that endogenous bile pigments such as bilirubin, particularly unconjugated bilirubin (UCB), have substantial anti-inflammatory and antioxidant properties that protect against oxidative stress-associated diseases such as CRC ,6,7,8,9.UGT1A1 gene results in the accumulation of UCB in the serum. Mild unconjugated hyperbilirubinemia, known as “Gilbert’s syndrome”, has a prevalence of 5–10% in Caucasians and is associated with a polymorphism of the 5′ end of the UGT1A1 gene promoter, a homozygous insertion of thymine–adenine (TA) pairs (genotype UGT1A1*28/*28) [Serum bilirubin is derived primarily from the degradation of hemoglobin and transported to the liver by binding to albumin. Within the hepatocytes, uridine diphosphate glucuronosyltransferase 1A1 (UGT1A1) is an enzyme that contributes to bilirubin glucuronidation . Bilirub*28/*28) . Finding*28/*28) ,23,24,25*28/*28) ,25. We rKORA is a regional research platform for population-based surveys. Between 1985 and 2001, four cross-sectional health surveys (S1 to S4) were performed, each comprising an independent random sample of all inhabitants of German nationality aged 25–74 years from the region of Augsburg, South Germany. Based on data on ethnicity from the S4 survey, the great majority of the study participants were of German ancestry (Caucasians), with some proportion (~14%) of migration, mainly from the eastern European region. The four cross-sectional surveys included 18,000 participants and serve as cohorts for long-term follow-up studies and as a pool for nested case-control and case-cohort studies. Follow-up activities include address inquiries for all participants , postal questionnaires focusing on chronic diseases, and complete follow-up studies with interviews and physical examinations. Detailed information about KORA has been published previously . The curData on sociodemographic characteristics and medical histories were assessed by standardized computer-aided interviews. The study participants provided information on physical activity during leisure time, alcohol consumption, and smoking status. The leisure-time activity was collected separately for summer and winter and was combined and categorized as active and inactive.Approval for the KORA study was obtained from the Ethics Committee of the Bavarian Medical Association and the Bavarian commissioner for data protection and privacy. All the participants provided written informed consent.Incident cancer cases in the KORA were identified by the active follow-up of study subjects and verified with cancer registry data. Our outcome of interest was incident first primary CRC (adenocarcinomas), irrespective of molecular subtype or stage, and was defined according to International Classification of Diseases, ICD-9  or ICD-10 (C18-C20). A total of 77 cases  were individually matched to 77 cancer-free (except non-melanoma skin cancer), living controls by participation in the same study , age at blood collection, and sex.Circulating UCB levels were measured in serum/plasma samples following a well-established protocol ,27 usingCase-control pairs were analyzed in the same plate to minimize batch-to-batch fluctuation. Bilirubin   acted as an external standard . One reference serum sample was assessed per analysis as an internal standard. The coefficient of variation (CV) between each plate was 3.76%.We decided a priori to perform all statistical analyses separately in men and women because of the well-established sex-differences in the levels of UCB, with average higher values in men ,29 and pThe baseline characteristics are presented as means and standard deviations for continuous variables or percentages (%) for categorical variables. Analyses were conducted using unconditional logistic regression models to estimate odds ratios (OR) and 95% confidence intervals (CI) for associations between log-transformed UCB levels (log-UCB), standardized per one-standard-deviation (one-SD) increment, and CRC risk with age, sex, and study adjustment in crude models and further adjustment for body mass index (BMI), height, alcohol consumption, physical activity (active and inactive), smoking status , and dietary patterns  , as well as hormone therapy  and menopausal status  in women in fully adjusted models. Potential effect modification by sex and menopausal status (among women) was tested by adding a multiplicative interaction term to the fully adjusted model.All statistical analyses and plots were performed using Stata SE14 . The significance testing was based on two-sided P-values of less than 0.05.We identified 77 participants who developed CRC during the follow-up period (between 7 (S4 and F4) and 26 (S2) years). heterogeneity values for differences by sex and menopausal status (among women) were equal to 0.4 and 0.7, respectively.Among the men, each one-SD increment in log-UCB was suggestively inversely associated with CRC risk in the crude model  as suggested by the lower UCB concentrations among the cases in the unadjusted analysis . HoweverWe investigated the potential etiological role of circulating UCB concentrations in CRC development using serum/plasma samples collected prior to cancer onset (diagnosis) among participants of the KORA cohort. In this population-based prospective cohort study among middle-aged men and women from southern Germany, we found little evidence for associations between circulating UCB concentrations and CRC risk. The confidence intervals around the point estimates were wide due to the limited number of cases (N = 77), and the findings did not reach formal statistical significance. However, the results of this study can be used in future meta-analyses on circulating bilirubin and the risk of CRC.The observed results of the fully adjusted models (non-significantly positive in men and non-significantly inverse in women) are congruent, in terms of the directions of associations, with our previously published work, where we also used a case-control study design, nested in the prospective EPIC cohort with 1386 cases and 1386 matched controls . In thisDespite the fact that in vitro work suggested that bilirubin might have antioxidant and anti-inflammatory effects, these inconsistencies in epidemiological studies are most likely attributable to differences in study design, sample size, or bilirubin behavior in human bodies. We used a prospective design with pre-diagnostic blood samples in the current analysis; however, the null results of the current study could be due to the lack of statistical power.Bilirubin is more than just the final product of heme catabolism. Today, it is considered to be a fundamental endogenously formed product in the blood, with antioxidant and anti-inflammatory properties . UCB hasThe strengths of our study include the case-control study design nested within a population-based cohort, the sizable follow-up period, and the direct measurements of serum UCB concentrations. We also had data for a number of demographic, lifestyle, and cancer-related variables to control for confounding. Our study was limited by, first, the low number of CRC cases, which reduced the statistical power of our analysis. Second, the storage of samples for prolonged periods of time  could have contributed to a degradation of UCB concentrations, which, however, should have affected the UCB concentrations of cases and controls equally.We found little evidence for associations of circulating bilirubin with CRC risk. However, the observed associations (non-significantly positive in men and non-significantly inverse in women) are congruent, in terms of direction, with previous evidence and may be used in future meta-analyses. Given the limited sample size in our study, we cannot exclude the possibility that bilirubin may play a role in CRC development, and further studies are required.


The high-temperature requirement factor A1 (HTRA1) gene located at 10q26 locus has been associated with age-related macular degenerative (AMD), with the significantly related polymorphism being , however, above association is not consistent. We investigated a comprehensive analysis to evaluate the correlations between rs11200638 polymorphism and AMD susceptibility thoroughly addressing this issue.An identification was covered from the PubMed and Wanfang databases until 27th Jan, 2020. Odds ratios (OR) with 95% confidence intervals (CI) were applied to evaluate the associations. After a thorough and meticulous search, 35 different articles  were retrieved.Individuals carrying A-allele or AA genotype may have an increased risk to be AMD disease. For example, there has a significantly increased relationship between rs11200638 polymorphism and AMD both for Asians  and Caucasians [OR (95%CI) = 2.63(2.29–3.02) for allelic contrast]. Moreover, a similar trend in the source of control was detected. To classify the type of AMD, increased association was also observed in both wet  and dry  AMD. Finally, based on the different genotyping methods, increased relationships were identified by sequencing, TaqMan, PCR-RFLP and RT-PCR.Our meta-analysis demonstrated that HTRA1 rs11200638 polymorphism may be related to the AMD development, especially about individuals carrying A-allele or AA genotype, who may be as identified targets to detect and intervene in advance. Further studies using Larger sample size studies, including information about gene-environment interactions will be necessary to carry out. In both developed and developing countries, age-related macular degeneration (AMD) is the main cause of vision loss in the elderly people , 2. By 2Age, race, family history, smoking and sun exposure are common risk factors , 8. AnotAs we all known, VEGF is contributed to the progression of wet AMD, because angiogenesis and the formation of vascular permeability can lead to fluid leakage in blood vessels, and eventually lead to loss of vision . Anti-VEhttps://www.ncbi.nlm.nih.gov/gene/5654).HTRA1 regulates the transforming growth factor-β (TGF-β), insulin-like growth factor, and its binding protein, which is considered as regulators for cell proliferation, angiogenesis and extracellular matrix deposition. Furthermore, the inhibition of TGF-β may result in the overexpression of HTRA1 gene in wet AMD   . The A-aIn view of the above, we are aware of the critial role of HTRA1 gene and its common rs11200638 polymorphism, and we conducted a comprehensive summary using meta-analysis methods, including 28 different publications (33 case-control studies) , 30–57.Relative studies from PubMed and Wanfang databases before 27th Jan, 2020 were searched. The keywords were “age-related macular degeneration,” “AMD,” “polymorphism or variant,” and “HTRA1 or high-temperature requirement factor A1.” Included criteria were according with as follows: (1) studies were focused on the correlation between AMD and rs11200638 polymorphism; (2) studies were all case-control and retrospective studies, and (3) the  genotypes both in case and control groups must be listed in Tables. Excluded criteria were consistent with as follows: (1) just case samples were shown in studies; (2) the numbers for genotypes did not shown in Tables, and (3) some duplications studies . After tTwo authors (Ying Liu and Dong Wei) independently screened all papers that according using above criteria. The basic information collected from each study contains the first author, year of publication, country source for corresponding authors, race type, Hardy-Weinberg equilibrium (HWE) for control group, detecting methods and AMD disease types (dry and wet AMD) ..Z-test   < 0.001, P < 0.001 in allelic contrast, Fig. P(heterogeneity) = 0.009, P < 0.001] and Caucasian populations  (Table P(heterogeneity) = 0.498, P < 0.001, Fig. Pheterogeneity = 0.073, P < 0.001, Fig. Pheterogeneity < 0.001, P = 0.002 about TaqMan; OR = 4.30, 95% CI = 2.51–7.35, Pheterogeneity = 0.073 about PCR-RFLP, OR = 3.84, 95% CI = 1.53–9.63, Pheterogeneity = 0.044 about MassARRAY MALDI-TOF, Fig. Pheterogeneity = 0.677 about sequencing, OR = 9.83, 95% CI = 5.18–18.65, Pheterogeneity = 0.817 about sequencing RT-PCR, Fig. Coming up, we all know that the frequency of A-allele in different races was not the same, so we tried to analysis the relationships by ethnicity subgroups in further, which indicated an incremental statistically association between this polymorphism and both in Asians [OR95% CI) = 2.512.22–2.83), , P; Begg’s test , Supplementary Figure Begg’s funnel plot and Egger’s test were applied to assess publication bias. At beginning, the funnel plots seemed asymmetrical in allele comparison for rs11200638 by Begg’s test, suggesting no publication bias was existed. Then, Egger’s test was applied to assess the funnel plot symmetry. As a result, no publication bias was observed [eg. allelic contrast, Egger’s test , it is necessary to study its etiology and mechanism, and then develop early diagnostic methods and effective treatments. Today, VEGF inhibitors have been widely regarded as effective drugs in clinical application for CNV (wet AMD) , 72, 73.In our analysis, we selected the HTRA1 gene that can regulate certain growth factors. The rs11200638 polymorphism in HTRA1 is the most common single nucleotide polymorphism (SNP) and has been received attention. However, Kanda et al.  demonstrMori et al. before all others reported the correlations for rs11200638 polymorphism and AMD . After tTwo previous meta-analysis , 77 abouOur current research is the comprehensive analysis about the associations between HTRA1 gene rs11200638 polymorphism and AMD, involving 8101 AMD samples and 7215 controls. Increased associations were found in the whole group, in Asian and Caucasian subgroups, source of control subgroup, and dry/wet sub-types of AMD, different genotyping methods , which means that A-allele or AA genotype is the risk factor for AMD, in other words, if individuals carry on this SNP from peripheral blood test, which may indicate that it is possible to increase the occurrence of AMD for them in present time or at some point in the future. Therefore, this polymorphism may be helpful in screening vulnerable populations for AMD in advance. In addition, the power of present study was 1.00, which suggested our conclusions were stable and convincing.ARMS2 and HTRA1 genes have a linkage disequilibrium, which is located nearby form 10q26 chromosome. ARMS2 rs10490924 was related to response to ranibizumab treatment among wet AMD patients [In addition, the online String website was used to make a forecast several potential and functional genes associated with HTRA1. As a result, ten genes were predicted. Among them, the highest score of association was ACAN (0.943), however, so far, no research has been reported between this gene and AMD and interaction between this gene and HTRA1. Future research should be payed attention to above information, which may be in favor of AMD early detection/prevention and intervention. In other partners, ARMS2 and CFH have been shown to associate with AMD. Both patients . CFH genpatients . As for patients , which sSome limitations should be declared. First of all, Mixed and African individuals should be paid more attention in future studies, which was vacant in present analysis. Second, analysis about gene-gene/gene-environment interactions should be added, because some specific environmental and lifestyle factors may influence associations between rs11200638 polymorphism and AMD . Third, vision is the most concerned-clinical indicator of AMD, future studies should include the value of the vision and analyze the relationships between rs11200638 polymorphism and the degree of visual impairment, which may help us to better detect disease progression.In conclusion, our present analysis demonstrated HTRA1 rs11200638 polymorphism may play a risk factor for the susceptibility of AMD, larger and more comprehensive studies should be performed in the future.Additional file 1: Table S1. Allele Frequency from 1000 Genomes Browser and present study.Additional file 2: Figure S1. A: Begg’s funnel plot for publication bias test . Each point represents a separate study for the indicated association. B: Egger’s publication bias plot .Additional file 3: Figure S2. Sensitivity analysis between HTRA1 gene rs11200638 polymorphism and AMD risk .


The overall pooled risk was 1.56  with moderate heterogeneity . Several differences among the included studies that may account for heterogeneity were investigated. Stratification for exposure indicator confirmed a positive association for studies assessing urine As. A slightly higher risk was detected pooling studies based in Asia rather than in North America. Stratification for GDM diagnostic criteria showed higher risks when diagnosis was made according to the Canadian Diabetes Association (CDA-SOGC) or World Health Organization (WHO) criteria, whereas a lower risk was observed when adopting the American Diabetes Association (ADA) criteria. These results provide additional evidence for a possible association between As exposure and GDM, although the data need to be interpreted with caution due to heterogeneity.Gestational diabetes mellitus (GDM) is a metabolic complication associated with adverse outcomes for mother and fetus. Arsenic (As) exposure has been suggested as a possible risk factor for its development. The aim of this meta-analysis was to provide a comprehensive overview of published evidence on the association between As and GDM. The systematic search from PubMed, MEDLINE, and Scopus was limited to full-length manuscripts published in peer-reviewed journals up to April 2020, identifying fifty articles. Ten studies met the inclusion criteria, nine for quantitative synthesis with a total of  Gestational diabetes mellitus (GDM), a common metabolic disease that affects up to 14% of pregnant women worldwide, is a glucose intolerance that develops during pregnancy and usually resolves after delivery . This coSeveral studies have found an association between GDM and As levels in maternal blood, urine and meconium, supporting the possibility that a high level of As exposure might predispose to maternal GDM. However, the data obtained so far are quite inconsistent ,31,32,33To offer an overview of the evidence available in the literature, we conducted a systematic review and meta-analysis on the plausible link between maternal As exposure and the risk of developing GDM.This systematic review and meta-analysis were performed according to the Preferred Reporting Item for Systematic Reviews and Meta-analysis (PRISMA) guidelines . The stuWe performed an advanced, systematic search of the online medical databases PubMed, Medline and Scopus using the following keywords: “arsenic” and “arsenic exposure” in combination with “gestational diabetes mellitus” or “diabetes in pregnancy”. Specific tools available in each database such as MeSH terms were used to optimize search output. Only manuscripts written in English and published in peer-reviewed journals up to April 2020 were included and duplicates were removed by using Endnote software version X9 . The potential eligibility of articles was first assessed by screening titles and abstracts. Then, full-text manuscripts were obtained and the final decision for inclusion was made after detailed examination of the articles. In order to identify any additional relevant citations, we also checked the reference lists of the eligible articles. The electronic search, the study selection and the eligibility for qualitative synthesis were independently assessed by two authors (R.V. and C.D). An independent author (N.S) assessed the eligibility for quantitative synthesis. Disagreements were resolved by discussion with a fourth reviewer (J.O.).The following predefined inclusion criteria were used to screen citations for eligibility: (i) exposure to As was assessed through an appropriate exposure indicator ; (ii) risk estimates were provided using odds ratio (OR) or relative risk (RR) with the corresponding 95% confidence interval (CI); (iii) study design limited to analytical studies ; (iv) outcome of interest was GDM and diagnosis of GDM was confirmed by a positive glucose challenge test  and/or a positive oral glucose tolerance test , according to the diagnostic criteria recommended by either the American Diabetes Association (ADA), the World Health Organization (WHO), the French National College of Obstetricians and Gynecologists , or the Canadian Diabetes Association and the Society of Obstetricians and Gynecologist of Canada (CDA-SOGC) ,37,38,39We excluded descriptive studies (case-report and case-series) and studies not reporting original results  as well as those dealing with the pathological condition of altered blood glucose levels not satisfying the diagnostic criteria for GDM ). Finally, studies were excluded from the quantitative synthesis  if a comparable estimation of effect size was not provided or in the sensitivity analyses.The following data from studies included in the quantitative synthesis were collected and tabulated by three independent reviewers  using a standardized data extraction form: (i) first author name, (ii) publication year, (iii) study country, (iv) study period, (v) study design, (vi) sample size, (vii) age and demographic data of the sample, (viii) number of cases, (ix) diagnostic method used to define cases, (x) exposure, (xi) exposure indicator , (xii) time of pregnancy when exposure was detected, (xiii) confounding variables in multivariate analysis, and (xiv) risk estimates with 95% CI.Two review authors (R.V. and C.D.) independently assessed the risk of bias by using the risk of bias tool for cohort studies developed by the Clarity Group  40]..40].We classified the possible sources of bias as definitely yes (low risk of bias), probably yes (moderate risk of bias), probably no (serious risk of bias), and definitely no , and then we assessed a comprehensive risk of bias judgment for each study included in our review.In the case of disagreements, resolution was achieved by discussion with a third reviewer (J.O.).Risk estimates with 95% CI were extracted by an independent reviewer (N.S.) from the original works. Almost all the studies included in the quantitative analysis presented odds ratios (ORs) and their 95% CIs. Relative risks (RRs) were converted in ORs . In stud2), the random-effect model suggested by DerSimonian and Laird was preferred for the quantitative synthesis of all included studies [p-value < 0.05 was interpreted as statistically significant. Sensitivity analyses were conducted by omitting one study at a time to explore the weight of each work in estimating pooled risks. Multivariate-adjusted risk estimates were transformed into log ORs and were pooled together using the generic inverse-variance approach as the model estimator with both fixed and random effect analysis. To incorporate the estimate of the pooled effect measure in the between-study variance  from the χ2 test indicated heterogeneity [Statistical heterogeneity of the intervention effects was assessed with χthan 75% . A low pogeneity . Potential publication bias was investigated by plotting the natural logarithm of the estimated OR (lnOR) against its standard error (SE). Asymmetry of the funnel plot was verified using the linear regression method proposed by Egger et al. .Subgroup analyses were performed following the guidelines suggested by Wang et al. . Risk esStatistical analysis was performed using RevMan software version 5.3 .The literature search identified 50 articles: among them, 10 met the inclusion criteria and the following characteristics were extracted ,31,32,33The flowchart of the systematic review is available in The plausible association between GDM and As exposure was assessed by the analysis of different human samples. Three studies collected blood samples, five papers analyzed urine samples, two evaluated arsenic concentration in home tap water, only one study measured As concentration in the meconium, and one paper in urine samples, home tap water, and toenails. Shapiro et al. used As in first trimester blood samples as an indicator of exposure, finding elevated odds of GDM in the highest quartile of As exposure in the adjusted analysis (adjusted odds ratio (aOR) = 3.7; 95% CI = 1.4, 9.6) . Similar results were obtained by Xia et al., who evaluated As levels in blood samples in the first and second trimester and cord blood, finding an association between GDM and As levels only for the fourth quartile of the first trimester samples . Stratified analyses showed the association was largely limited to normal maternal age  and normal weight women  . 2)  and high pre-pregnancy BMI (≥24 kg/m2)  in the second tertile [The cohort study conducted by Wang et al., which evaluated blood samples taken the day after delivery, showed an increased risk of GDM for the second tertile . This risk was even higher among women with low pre-pregnancy BMI . However, a significant association between arsenic concentration and risk of GDM was found only in the single metal model (p = 0.019) without any validation in the multiple-metals model analysis  [Levels and GDM (anadium) .Ashley-Martin et al. analyzed urinary metabolites (DMA and arsenobetaine) of As, stratifying results for urinary specific gravidity. They found a significantly increased risk of GDM  in women with DMA concentration higher than 3.52 μg As/L (third tertile). Interestingly, the aOR was even higher when the analysis was restricted to women carrying male infants  .The study conducted by Khan et al. demonstrated that As level in urine might predict the likelihood of having GDM . However2)  [The findings from Farzan et al. found a close relationship between As exposure via home well water and risk of GDM: each 5 μg/L increase in As concentration in home well water was associated with a 10% increased odd of GDM . This association was largely limited to obese women  . 2) . The French correlational study carried out by Marie and colleagues  providedOnly one study investigated the link between GDM and As exposure in meconium, finding a higher concentration of the metal in studied cases when compared to controls. Arsenic levels were positively associated with maternal GDM with aORs of 3.28 , 3.35  and 5.25  for the second, third, and fourth quartiles, respectively . One of the included studies investigated the association between As exposure and the risk of GDM measuring As concentrations in toenails. A positive and statistically significant association was observed: each 100% increase in toenail As was associated with a nearly four-fold increased risk of GDM (aOR = 4.5), despite the wide confidence interval  . The forest plot of the meta-analysis including all studies for As exposure and the risk of GDM is reported in n = 9) the pooled OR calculated according to the random effect model was 1.56 , with obvious moderate heterogeneity  and slightly high publication bias  [p for effect = 0.0003). The meta-analysis performed using the fixed effect model showed quite similar results . Sensitivity analysis conducted by omitting one study at time (n = 8) revealed that the result of the pooled analysis was quite robust. For all the included studies ( = 0.02) ,31,32,332 = 1.86; p = 0.17; I2% = 46) and high pooled risk  when combining data from cross-sectional studies rather than when pooling data from the cohort studies . The results of the different meta-analyses performed are reported in Further analyses were performed to investigate the possible causes of heterogeneity, stratifying studies according to exposure indicator, study country, and diagnostic criteria for GDM.2 = 4.20; p = 0.24; I2% = 29), moderate heterogeneity when pooling studies measuring tap water As , and high heterogeneity when studies based on blood As were combined together . The pooled effect estimates according to stratification by exposure indicator carried quite similar results for urine and blood As , whereas a minor association was found for tap water As . We found low heterogeneity when combining studies assessing urine As  and in Asia . The pooled risk estimate was slightly higher for studies based in Asia , rather than in North America . For studies based in North America, a sensitivity analysis was conducted by omitting Shapiro et al., since redundancy of data between Shapiro et al. and Ashley-Martin et al. could not be excluded [When combining data from different study countries, we found a similar high heterogeneity for studies conducted in North America , low heterogeneity when pooling studies adopting ADA diagnostic criteria , and moderate heterogeneity when studies defining cases according to WHO diagnostic criteria were combined together . For studies where diagnosis of GDM was based on ADA criteria, a sensitivity analysis was performed by omitting Farzan et al. [Stratification by diagnostic criteria of GDM showed higher pooled risk estimates when diagnosis of the disease was made according to CDA-SOGC criteria  or WHO criteria  rather than with ADA criteria . We found no heterogeneity when combining studies where GDM diagnosis was established with CDA-SOGC criteria . Three studies reported statistical support to the relationship between As exposure and risk of GDM using total As in blood as the exposure variable ,25,26. T2% = 29), suggesting a possible more accurate assessment of As exposure when using urine As as the exposure indicator. Five studies used urine samples in order to assess the association between exposure to As and GDM ,29,30,312 % = 60). Water As level represents a valid exposure measure for inorganic As if it is the primary source of exposure and individual water intake levels are known. However, it might underestimate the exposure among people with high inorganic As intake from foods , leading to altered exposure assessment [The association between exposure to As in tap water and GDM was assessed in two studies, which reported a significant association ,33. In lsessment ,57.Arsenic levels in maternal toenail samples and fetal meconium were also analyzed by two different studies that found a significant association ,32. Toen2% = 77) or when pooling Asian studies together (I2% = 76). Moreover, a substantial reduction in heterogeneity among Asian studies (I2% = 58) was observed when the analysis was performed by omitting the study by Peng et al. [Grouping of the studies by study country did not reduce heterogeneity either for studies conducted in North America . Nevertheless, both studies included according to this stratification [2% = 44), while the main source of heterogeneity came from studies based on WHO diagnostic criteria (I2% = 70). This finding could be explained by a different definition of cases in the study based in Chile [Combining studies according to the different criteria adopted for GDM diagnosis, we found no heterogeneity among studies based on CDA-SOGC diagnostic criteria  , we inclThe major limitation of this meta-analysis is the strong heterogeneity and degree of inconsistency existing between the nine individual risk estimates. Several differences between the included studies that may account for this heterogeneity were analyzed including study design, exposure indicator, study country, and diagnostic criteria of GDM. The random model estimator analysis did not substantially change the risk estimates and no reduction in heterogeneity among the included studies was observed when adopting this model. As a matter of fact, a fixed meta-analysis has natural complements that provide heterogeneity , thus measures of heterogeneity should not be used to determine if this model could be appropriate ,68. IndeIn summary, the results of this systematic review and meta-analysis provide additional evidence for a possible association between As exposure and the risk of GDM. To improve and confirm the available data, future study designs might benefit from the inclusion of standardized methods with more sensitive limits of exposure detection in order to evaluate the effects of inorganic and organic As on glucose homeostasis during early pregnancy, hence prior to GDM diagnosis. Additionally, as controversy still surrounds the diagnosis of GDM, a universally endorsed diagnostic criteria could help in confirming the potential role of As in contributing to the onset of this disease, hopefully implying new prevention strategies to reduce the burden of GDM worldwide.


BackgroundBeta (β)-thalassemia major patients frequently suffer from many vascular problems. Thrombophilia is a blood disorder that comprises imbalances in the blood coagulating factor due to ecological and hereditary components. Previous evidence shows that thrombosis is the commonest risk in beta-thalassemia patients. Several studies have examined that MTHFR C677T, prothrombin G20210A (PT G20210A), and Factor V Leiden G1691A (FVL G1691A) polymorphism play a crucial role in the development of β-thalassemia major, yet the result was questionable and uncertain. Therefore, in this study, we executed the correlation between these gene polymorphisms with β-thalassemia major patients.MethodsSuitable keywords were used to search related articles in PubMed, Google Scholar, and Web of Science. In this random-effects meta-analysis, we analyzed the odds ratio (OR) for the estimation of risk.Results2=61%) and allele C vs T (I2=71%) of MTHFR and genotype GG vs GA (I2=95%), GG vs GA+AA (I2=95%), GA vs GG+AA (I2=95%), and allele G vs A (I2=93%) of FVL G1691A.A total of nine research articles with 645 β-thalassemia major patients and 989 healthy controls were incorporated in this meta-analysis. The pooled OR was assessed in MTHFR C677T, PT G20210A, and FVL G1691A polymorphism. This random-effects meta-analysis demonstrated that MTHFR C677T, PT G20210A, and FVL G1691A gene polymorphism did not significantly associate with β-thalassemia major. Moreover, the heterogeneity was significantly found in genotype CC vs CT+TT C677T  gene polymorphism are not a risk factor for β-thalassemia major. Beta (β)-thalassemia major is a single genetic disease affected by mutations in β-globin genes. These mutation causes a complete deficiency or reduces the formation of β-globin chains . This diHighly thromboembolic events have been associated with β-thalassemia, which leads to the thrombosis of arterial or venous blood -6. VarioMTHFR C667T gene polymorphism altered the function of enzymes and increased the risk of thrombosis . A previVarious epidemiological studies have explored the associations of the MTHFR C677T, PT G20210A, and FVL G1691A gene polymorphism with β-thalassemia. Still, the outcomes of the studies were conflicting and indecisive. In light of the above evidence, we executed this meta-analysis of accessible articles connecting MTHFR C677T, PT G20210A, and FVL G1691A gene polymorphism to the risk of β- thalassemia.Setting and study designThis meta-analysis study was done in the Center for Advance Research (CFAR), Cytogenetics Lab, King George’s Medical University, Lucknow.Identification and methods: search and selection of studiesβ-thalassemia related articles were independently recognized by investigators from PubMed, Google, and Web of Science. The search terms were as follows:  AND . Moreover, the references of all articles and reviews were manually searched for extra appropriate studies.Results were restricted to MTHFR C677T, PT G20210A, and FVL G1691A gene polymorphism and β-thalassemia major in humans. A total of 780 articles were identified in PubMed, Google, and Web of Science. Inclusion criteria for the selection and inclusion of studies were: i) Case‑control study of β-thalassemia patients and controls ; (2) MTHFR C677T, PT G20210A, and FVL G1691A gene polymorphism; (3) Published articles; (4) Full article in English; (5) Studies with the expression of MTHFR C677T, PT G20210A, and FVL G1691A gene, and studies on animals or in the lab were excluded. The association of MTHFR (C677T), PT (G20210A), and FVL (G1691A) gene with β-thalassemia and control was the main concern for the selection of studies.Data collectionData were extracted by authors individually from articles text, figures, or tables in each published article. The above-mentioned details were collected from individual studies, i.e., types of SNP, the technique for polymorphism, and the number of patients and controls Table .Data synthesisThe data were obtained from every study: title of the article, name of the principal author’s, journal name, date of publication, name of nation, and strategy utilized for genotyping.Statistical analysisRevMan 5.3 (Review Manager version 5.3) of the Cochrane Collaboration, London, United Kingdom, was used for statistical interpretation. The odds ratio (OR) was used for risk assessment. In addition, the heterogeneity among studies was estimated by the Q test with p-values <0.05 and >50% I2 was considered significant. Initially, the associations-based studies were analyzed; after that, the subgroup examinations by SNP type were accomplished to find the SNP type-specific impacts.Search resultsThe search outcomes of published articles have been shown in Figure MTHFR C677T gene polymorphism and the risk of β-thalassemiaFor MTHFR C677T polymorphism, a total of eight studies, including 612 β-thalassemia patients and 871 healthy controls were examined Table .Prothrombin G20210A gene polymorphism and the risk of β-thalassemiaFor prothrombin G20210A polymorphism, a total of six studies, including 521 β-thalassemia patients and 863 healthy controls were examined Table .Prothrombin G20210A gene polymorphism and the risk of β-thalassemiaFor prothrombin G20210A polymorphism, a total of six studies, including 521 β-thalassemia patients and 863 healthy controls were examined Table .Factor V Leiden G1691A gene polymorphism and the risk of β-thalassemiaFor Factor V Leiden G1691A polymorphism, a total of four studies, including 484 β-thalassemia patients and 681 healthy controls, were examined Table . This raMeta-analysis is a potent analyzing tool for accumulative data with low and small power studies. In this study, we explore the correlation of MTHFR C677T, PT G20210A, and FVL G1691A gene polymorphism with β-thalassemia major patients. Here, we meta-analyzed the published information about the polymorphism of MTHFR C677T of 612 β-Thalassemia major patients from eight published studies and, particularly, the changes in allele and genotype occurrences of MTHFR C677T gene polymorphism in patients as compared with controls. We also meta-analyzed the published data about the polymorphism of PT G20210A and FVL G1691A of 521 and 484 β-thalassemia major patients from six and four studies, respectively, and compared the genotype and allele frequencies of PT G20210A and FVL G1691A polymorphism in patients as compared with controls.In this random-effects meta-analysis, we did not find an association between MTHFR C677T, PT G20210A, and FVL G1691A gene polymorphism with β-Thalassemia major patients as compared to controls. Overall, it can be concluded from general and subgroup analyses that the MTHFR C677T, PT G20210A, and FVL G1691A gene polymorphisms may not play an important role in the pathogenesis of a thromboembolic event. Still, there is not a sufficient amount of relevant studies to give a safe and good assumption. In the future, however, more well-designed studies with a bigger sample size and metacentric studies will be necessary to validate the current outcomes.In meta-analyses, heterogeneity is one of the significant factors. In this study, heterogeneity was significantly observed in genotype CC vs CT+TT (I2=61%) and allele C vs T (I2=71%) of MTHFR C677T polymorphism and genotype GG vs GA (I2=95%), GG vs GA+AA (I2=95%), GA vs GG+AA (I2=95%), and allele G vs A (I2=93%) FVL G1691A gene polymorphism. However, heterogeneity was not significantly found in other genotypes of MTHFR C677T and FVL G1691A and the genotype and alleles of PT G20210A gene polymorphism.A previous study reported that the T allele  frequencies of MTHFR C677T polymorphism were 21.5% and 21% in β-Thalassemia major and healthy controls, respectively. The frequency of MTHFR C677T mutations was marginally greater but not significantly different between β-thalassemia major patients and controls ,27. The The mutation of the MTHFR C677T gene was not statistically correlated with an increased risk of thrombophilia in patients with β-thalassemia major ,22. ContWhen interpreting the meta-analysis outcomes, various limitations should be considered. First of all, the appropriate studies were limited in number. To limit the conceivable outcomes of distribution bias, published articles were incorporated. Thereafter, the present results were based on unadjusted appraisals. Finally, our analyses did not reflect the possibility of linkage disequilibrium and interaction between gene-gene or SNP-SNP.The results of the present meta-analysis suggest that the genotype and allele frequencies of MTHFR C677T, prothrombin G20210A (PT G20210A), and Factor V Leiden G1691A (FVL G1691A) gene polymorphism were not significantly correlated with the risk for the development of β-thalassemia major. However, the outcomes of this meta-analysis were based on mutation (single gene polymorphism), and significant heterogeneity was likewise identified. Therefore, the outcomes of the current study should be interpreted and need further well-designed studies.


This paper describes data from a systematic review and meta-analysis [1] conducted to identify and evaluate published peer reviewed evidence on the association between perineal use of talc powder and risk of ovarian cancer. These data were collected from multiple electronic bibliographic databases, as well as from grey literature sources, without applying time, language or other filters. A meta-analysis was conducted to quantitatively assess the ovarian cancer risk in relation to talc use and other potential risk factors. A robust, multi-step search and screening of evidence was conducted to identify original, peer-reviewed evidence from human subjects. Extracted data include detailed population demographics and risk of ovarian cancer in relation to a wide range of possible risk factors. Data extracted from these studies are summarized in Table 1 in the main review, and detailed in We conducted meta-analyses through pooling and analyzing the reported risk estimates, while ensuring the inclusion of data-rich studies and avoiding the inclusion of multiple studies reporting on the same population. Data extracted from these studies are summarized in Table 2 in the main review, and detailed in 2We conducted a robust systematic search of 8 electronic bibliographic databases and other grey literature sources for original human studies examining the association of perineal use of talc powder and the risk of ovarian cancer. The search was conducted according to the PRISMA  guidelines, and following the specific guidance provided by the Cochrane Collaboration . SearcheA robust search strategy was designed and conducted by two reviewers (MT and NF), followed by automated then manual removal of duplicate studies. Identified studies were independently screened and assessed by the two reviewers using Distiller SR software . A totalQuality of the included studies was also assessed independently using the Newcastle-Ottawa Scale  based onhttps://ars.els-cdn.com/content/image/1-s2.0-S0890623818306373-mmc1.pdf].A summary of the systematic review strategy is provided in the I2, a Chi-Squared statistic, to assess heterogeneity among studies combined in the meta-analysis [In conducting the meta-analysis, we considered the reported maximally-adjusted risk estimates  to be practically equivalent, based on the rarity of the endpoint of interest, ovarian cancer. We pooled these estimates using random effects model with inverse variance weighting to obtain an overall risk estimate and associated 95% confidence interval [The initial analysis involved an evaluation of ovarian cancer risk in relation to genital/perineal use of talc powder (ever vs never use). Sixteen additional subgroup analyses were performed to assess the effects of duration and frequency of talc use, tumor histology, tumor behavior, and the possible effect of menopausal state, hormone use, and pelvic surgery. Additionally, we assessed possible effects of the study design, type of study controls, Newcastle Ottawa Scale (NOS) quality score , and pub


Background: Several epidemiological studies have been performed to evaluate the association of dietary intake of vitamin C-oriented foods (DIVCF) with risk of fracture and bone mineral density (BMD) loss, but the results remain controversial. Therefore, we conducted a systematic meta-analysis to assess this correlation.Methods: We searched EmBase, PubMed, Web of Science, and the Chinese database CNKI for relevant articles published up to August 2019. Pooled relative risks (RRs) with 95% confidence intervals (CIs) were calculated using the random- or fixed-effects model. Discrepancies were resolved by consultation with a third expert.Results: A total of 13 eligible articles (including 17 studies) with 19,484 subjects were identified for the present meta-analysis. The pooled RR of hip fracture for the highest vs. lowest category was 0.66  for DIVCF, i.e., people with a greater frequency of Vitamin C uptake had a 34%  lower prevalence of hip fracture. In subgroup analyses stratified by study design, gender, and age, the negative associations were statistically significant. Furthermore, the statistical analysis of the association between DIVCF and risk of osteoporosis , BMD at the lumbar spine , and BMD at the femoral neck  showed beneficial effects of DIVCF.Conclusion: Our meta-analysis indicates that DIVCF is negatively associated with the risk of hip fracture, osteoporosis, and BMD loss, suggesting that DIVCF decreases the risk of hip fracture, osteoporosis, and BMD loss. Osteoporosis is a systemic skeletal disease, which is common in postmenopausal women and is characterized by an increased risk of bone fragility and a decrease in bone mass . Bone hoIncreasing evidence demonstrates that osteoporosis is affected by genetic factors, low body mass (weight), and lifestyle factors, such as coffee intake, alcohol consumption, soft drink intake, and dietary intake of vitamin C-oriented foods (DIVCF) . Diet anAlthough a series of epidemiological studies have been performed to evaluate the association between DIVCF and the risk of osteoporosis –22, the The PRISMA  guidelines and others were followed in the current analysis , 24.A systemic search was performed for potential articles in four databases, i.e., EmBase, PubMed, Web of Science, and the Chinese database CNKI, published in or before August 2019. We used the search terms “Vitamin C,” “ascorbic acid,” “ascorbate,” and “acid ascorbic” in combination with “osteoporosis” or “fracture.” The reference lists from the articles included were manually screened for undetected relevant studies. The final results of the literature search were updated on August 31, 2019. The detailed steps of the literature search are shown in Articles were included in this review if they met the following criteria: (1) observational study published as an original study; (2) exposure of interest was defined as DIVCF, i.e., dietary intake of foods that are rich in Vitamin C; (3) outcome of interest was defined as osteoporosis, fracture, and/or BMD loss in the subjects; and (4) study included estimates of relative risks (RRs), comparing the highest DIVCF category with the lowest, with corresponding 95% confidence intervals (CIs) for the associations or other information that could be used for inference. Furthermore, the following exclusion criteria applied: (1) studies describing animal experiments, review papers, and mechanistic studies; (2) studies lacking specified data of osteoporosis in relation to low bone mass; and (3) articles that only contained an abstract. All identified articles were reviewed independently by two researchers, and discrepancies were discussed with and resolved by a third expert.Two researchers extracted data independently, following the standardized norms for literature collection. Any discrepancies were resolved by consulting with a third expert. The following information was extracted from the articles: name of the first author, publication year, country where the study was conducted, baseline age, gender, study design, sample size and number of cases, assessment approach of DIVCF, assessment approach of osteoporosis, the measurement location of osteoporosis, and adjusted RR with 95% CI of osteoporosis for DIVCF consumption. The Newcastle–Ottawa Scale (NOS) was used to assess the quality of case–control and cohort studies . The AgeI2 ≥ 50%) was observed; otherwise, the fixed-effects model (FEM) was used. The I-squared (I2) statistic and chi-square test were adopted to explore the possible heterogeneity of the studies. Potential publication bias was inspected by funnel plots. Subgroup analyses were performed by study type, gender, and age to assess potential influencing factors. All statistical analyses were performed using the statistical software package STATA SE, version 14.1 .The risk of osteoporosis, fracture, and BMD loss was estimated, comparing the highest DIVCF category to the lowest. Pooled RRs with 95% CIs were calculated using the random-effects model (REM) if moderate or high heterogeneity , we conducted subgroup analyses based on study type , gender, and age  (Eleven studies from seven articles –16 reporubjects) –5.Regarding the studies assessing the correlation between DIVCF and risk of hip fracture, two articles (including four case–control studies) , 16 repoSeven articles with 11 studies) –16 repor studies The RR of hip fracture, from three articles (with four studies) , 13, 15 Funnel plots were used to evaluate the potential publication bias, with RRs generated from the articles concerning DIVCF and the risk of hip fracture. In the absence of publication bias, most of the formulated points are symmetrically placed around the vertical line given by the pooled RRs. No evidence of publication bias was found in the evaluation of DIVCF and the risk of hip fracture .Four studies –19, 21 rr, 0.15; 95% CI, 0.09–0.23). Furthermore, four studies   .r, 0.15; 95% CI, 0.09–0.23), and BMD at the femoral neck  showed high DIVCF has beneficial effects. Our meta-analysis indicates that DIVCF is negatively associated with the risk of hip fracture, osteoporosis, and BMD loss, suggesting that people should consume more Vitamin C to decrease the risk of hip fracture, osteoporosis, and BMD loss.Our meta-analysis included 13 eligible articles (including 17 studies) with 19,484 subjects. The pooled RR of hip fracture for the highest DIVCF category vs. the lowest category was 0.66 , i.e., people with a high frequency of Vitamin C-oriented food intake had a 34%  lower prevalence of hip fracture. In our subgroup analyses stratified by study design, gender, and age, the negative associations were found to be statistically significant. Furthermore, the analysis of DIVCF and risk of osteoporosis , BMD at the lumbar spine  and at the femoral neck   .Although a series of studies on Vitamin C and the risk of osteoporosis have been published, including controlled trials, case series, and case reports, to the best of our knowledge, no systematic meta-analyses have been published focusing on the correlation between DIVCF and osteoporosis, hip fracture, and BMD loss.Several biological mechanisms may underlie the impact of DIVCF on the risk of osteoporosis. First, DIVCF includes not only Vitamin C but also Vitamin A, Vitamin B1, Vitamin B2, Vitamin B3, iron, and potassium intake. The positive effects of these vitamins and minerals on bone health have been well-established . OsteopoA positive link was found between DIVCF and a reduced risk of osteoporosis, fracture, and BMD loss. Several observations support the use of this information in clinical settings. First, our meta-analysis included a large number of subjects, reducing the sampling error to a large extent. Second, large efforts have been made to control the interference of possible confounding factors , resulting from the adjusted RR being fully extracted. Third, similar correlations were observed after performing subgroup analyses stratified by study type, gender, and age, confirming that the results are credible and robust.Nevertheless, our study has several limitations. One of the key points is that only DIVCF was taken into account , while many people with high DIVCF may in general eat a healthy diet, so an analysis of other healthy diets may give similar results. Therefore, it is necessary to verify that DIVCF is the active component of the diet. Also, it would be helpful to include intake of other dietary components (such as fruit and vegetable) for further study. Furthermore, Vitamin D plays a large role in the maintenance of BMD; its influence on the development of secondary hyperparathyroidism and osteomalacia, states where BMD is reduced reversibly, has been reported. Only a few studies in our meta-analysis adjusted for Vitamin D intake, though it is of great importance in the majority of nutritional studies , 18–20. Another important point is that most of the articles we identified were cross-sectional or case–control studies; larger prospective cohort studies are warranted to verify the results. Although we have adjusted for most of the confounders the articles included, other confounders may still have existed. Moreover, the confounders that were adjusted for were different in each research; this may certainly have affected the results. Moreover, the approach of osteoporosis assessment differed between articles, which could influence the observed correlation. Because of the limited number of included articles, the effects of an inconsistent assessment approach on the strength of the correlation could not be further explored. Finally, due to the limited available data, we could not assess the dose–response association between DIVCF and the risk of osteoporosis.Between-study heterogeneity is identified in most meta-analyses, and it is of importance to explore the sources of heterogeneity between studies. In our meta-analysis, there was high heterogeneity in the correlation between DIVCF and risk of osteoporosis. The factors contributing to such heterogeneity are complex. First, differences in the definition, classification, and assessment of DIVCF were found, which could result in heterogeneity. Second, the variables adjusted for  were not consistent between studies. In addition, because of the limited sample size, the evaluation of publication bias calls for further research.Our meta-analysis indicates that DIVCF is negatively associated with the risk of osteoporosis, hip fracture, and BMD loss and suggests that people should consume more DIVCF to reduce these risks.JL and Z-PL conceived and designed the study. The literature searches, study selection, critical appraisal, data extraction, and contacting authors of included studies for additional information were carried out by L-FZ, M-HL, XX, Z-TG, DZ, J-LZ, J-TL, and J-KP. G-HL, JY, XW, H-YC, H-TH, QL, Y-HH, J-HL, S-RH, MW, and L-FZ conducted the interpretation and analysis of the relevant data. L-FZ, W-YY, DG, W-XL, QW, and JL drafted the paper. JL, A-HO, and L-FZ further revised the text. The final version of this article was rechecked and approved by L-FZ, M-HL, G-HL, M-HL, W-YY, XX, XW, JY, DG, H-YC, J-KP, H-TH, QL, Z-TG, Y-HH, DZ, J-LZ, S-RH, MW, J-TL, J-HL, W-XL, A-HO, QW, Z-PL, and JL.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.


MBL2 polymorphisms and PTB.Mannose-binding lectin (MBL2) is considered to play a role in the human innate immune response to tuberculosis (TB) infections, and 4 common single nucleotide polymorphisms (SNPs) may be associated with pulmonary tuberculosis (PTB) risk. To examine these potential associations, we performed a comprehensive analysis to assess the relationships between The PubMed, Embase, and SinoMed databases were searched for articles published prior to June 13, 2019. Odds ratios with 95% confidence intervals were calculated to evaluate the strength of the relationships.MBL2 on PTB. A positive association between rs11003125 and PTB risk was observed in the hospital-based subgroup. Moreover, for the combined polymorphism and PTB risk, positive associations were detected not only in the total population but also in those with Asian origins across all source of control subgroups. No associations were found for rs7096206 or rs7095891.There were 37 case-control studies examining the effects of the four SNPs in MBL2 may be associated with susceptibility to PTB.Our current study indicated that several SNPs in  Mycobacterium tuberculosis go on to develop TB [interferon-gamma (IFNG), vitamin D receptor (VDR), solute carrier family 11a member 1 , and mannose-binding lectin (MBL2) [Tuberculosis (TB) is a global public health issue that poses serious threats to human health. It has been estimated that 1/3 of the world’s population may be infected with tubercle bacilli, but only 1/10 of individuals infected with velop TB –6. In favelop TB , 8. Prevn (MBL2) –12.MBL2 gene, which is a member of the complement system, has been hypothesized to play a dual role in the innate immune response to infections by activating the classical lectin pathway and by phagocytosis [MBL2 on chromosome 10 , which comprises four exons. MBL2 is reported to have several genetic polymorphisms that are commonly associated with MBL serum levels. Three point substitutions, located at codons 52, 54, and 57 in exon 1, are supposed to disrupt the assembly of MBL trimers or accelerate the degradation of the protein, thereby causing a decrease in the functional activity of MBL in the serum. These mutations are frequently referred to as variants D, B, and C, respectively, and they are collectively known as O, while A is the wild type. In addition, three other point substitutions have been reported in the nonstructural region: two at positions − 550 (H/L variants) and − 221 (X/Y variants) in the promoter region and one point mutation at position − 4 (P/Q variants) in the 5′-untranslated (UTR) region [The ocytosis , 14. MBLocytosis . Human M) region –18.MBL2 gene variations and pulmonary TB (PTB) risk [Many epidemiologic studies, including meta-analyses, suggest that there are relationships between TB) risk , 19–47. MBL2”. A total of 163 papers were identified, 30 of which were consistent with our criteria. The inclusion criteria for papers were as follows: (i) examined the relationship between PTB susceptibility and MBL variations, (ii) case-control study, and (iii) contained a complete number of genotypes (MM + MW + WW) among cases and controls. The exclusion criteria were as follows: (i) no control group, (ii) incomplete genotype frequency data, (iii) duplicate publication, and (iv) controls did not meet the Hardy-Weinberg equilibrium (HWE) standards.The PubMed, Embase, and SinoMed databases were searched for articles published prior to June 13, 2019, using the keywords “tuberculosis,” “TB,” “polymorphism,” and “mannose binding lectin 2 or The essential data are listed as follows: first author name, publication year, original country, race, total samples of case/control, each genotype in cases/controls, source of control and genotype methods. Race was classified as Caucasian, Asian, African, or mixed. The source of control subgroups included population-based (PB) and hospital-based (HB) subgroups. The type of TB included total TB, PTB, and EPTB.The NOS was used to assess the quality of each study and to assess the various aspects of the methodology, including the selection of cases, the comparability of groups and the determination of exposure. The total score on the NOS ranges from 0 to 9 stars. Studies with scores greater than 7 are considered high-quality studies .MBL2 and PTB risk based on the genotype frequency of the case and control groups. The Z-test was used to determine the statistical significance of the correlations. The heterogeneity between the studies was evaluated using a Q-test based on the χ2 method. In the Q-test, a P value greater than 0.05 indicates that there is a lack of heterogeneity between the studies. Because the Q-statistic does not reveal the statistical significance of the heterogeneity, the I2 test was applied to better assess the extent of heterogeneity. As a guide, I2 values are divided into three categories , corresponding to low risk, medium risk, and high risk, respectively [P ≤ 0.05 or I2 ≥ 50%, a random effects model was adopted; otherwise, a fixed effects model was used [MBL2 and PTB risk by testing the allelic contrast , heterozygote comparison , homozygote comparison , recessive genetic model  and dominant genetic model . Sensitivity analysis was applied to assess the stability of the results. The HWE was evaluated by Pearson’s χ2 test, and P = 0.05 was considered statistically significant [We used 95% CIs to measure the correlation between SNPs in ectively . If P ≤ ≤25%, 25–%, ≥50%, was used . We accenificant . Publicanificant . All staMBL2 were derived from the literature in Table Methods for genotyping the SNPs in MBL2 and TB susceptibility were included . After filtering out studies that met our exclusion criteria, 9 different case-control studies were included for rs7096206, 6 for rs11003125, 7 for rs7095891, and 15 for the A/O SNP : 0.755, P: 0.017, in the allelic contrast; OR: 1.91, 95% CI: 1.07–3.40, P(heterogeneity): 0.571, P: 0.029, in the homozygous comparison model : 0.633, P: 0.033 in the dominant genetic model).The associations between the 4 SNPs in P < 0.001 for heterogeneity, P: 0.018; dominant genetic model: OR: 1.41, 95% CI: 1.06–1.86, P < 0.001 for heterogeneity, P: 0.017; allelic contrast: OR: 1.33, 95% CI: 1.05–1.70, P < 0.001 for heterogeneity, P: 0.020, Fig. P: 0.001 for heterogeneity, P: 0.017, Fig. P: 0.005 for heterogeneity, P: 0.044; dominant genetic model: OR: 1.28, 95% CI: 1.04–1.57, P: 0.003 for heterogeneity, P: 0.021, Fig. P: 0.283 for heterogeneity, P: 0.001, Fig. P < 0.001 for heterogeneity, P: 0.037, Fig. For the A/O combined SNP (AA/AO/OO) polymorphism, the O allele had a positive association with PTB risk in the total sample . In recent decades, the effect of host susceptibility genes on TB has been increasingly recognized along with the development of genetic susceptibility. However, recent studies on the associations between SNPs in  bacilli , 54, ind alleles , 55. TheMBL2 polymorphisms and susceptibility to TB; however, each meta-analysis has its own conclusion and merits. Cao et al. analyzed 22 studies to assess the effect of MBL2 polymorphisms on TB risk. The rs1800451 polymorphism was associated with decreased TB risk in both the total sample and in some ethnic groups; in addition, A/O, rs7096206 and rs1800450 were likely only related to risk in some ethnic groups [MBL2 structural gene variants  with TB susceptibility. They did not find a significant association between the MBL2 genotype and PTB infection. By contrast, a meta-analysis of four studies examining MBL levels and susceptibility to TB found a significant association of high MBL levels with susceptibility to TB, although increased serum MBL levels due to the acute-phase reaction could not be ruled out. In addition, Areeshi et al. [MBL2 polymorphisms and TB risk in Asians. The results indicated a protective role of alleles B and C in TB infection. Finally, Shi et al. [MBL2 codon 54 B allele had an increased risk of TB compared with AA homozygotes, whereas rs7095891 was possibly not associated with TB risk in Chinese.Several meta-analyses have focused on the relationships between c groups . The anac groups . Their sc groups  examinedi et al.  found a i et al.  indicateMBL2 variants and PTB susceptibility. This analysis involved approximately 12,052 patients with PTB and 13,905 healthy samples. The most important finding of our study was that the rs11003125 L-allele and the A/O combined SNP were risk factors for PTB susceptibility in the HB subgroup, which was similar to findings from a previous meta-analysis. The O allele was also a risk factor for PTB in the Asian and PB subgroups. The aforementioned conclusions were novel concepts that have not been found in previously published meta-analyses.To our knowledge, the current study is an updated systematic analysis exploring the relationships between MBL2 variants on susceptibility to TB. One possible explanation for this effect is that different polymorphisms may have different effects on gene function, resulting in changes in PTB susceptibility. Second, the complex interaction between several genetic and environmental factors may involve the development of PTB. We think these conflicting results among studies and different populations suggests linkage disequilibrium with other nearby genes (e.g., surfactant proteins A1, A2 and D [MBL2 variants and PTB. Third, it is now widely accepted that differences in ethnicities between cases and control measures may be a source of confusion in the compilation of studies. Fourth, research with “negative” results takes longer to publish due to the time-lag bias, and positive research results are published much faster. Fifth, small studies of with “negative” results have never been published, and small studies of similar quality with “positive” results will also be shown in the literature [The above contradictory results from previous meta-analyses further emphasize the controversy about the effect of A2 and D  previousterature –63. SixtMBL2 gene. Third, other factors such as age, sex, smoking, familial history, disease stage, specific environmental factors and lifestyles should be included. Fourth, only one article [MBL2 structural gene variants and susceptibility to PTB. Further studies should aim to overcome these limitations.Some limitations in our study should be noted. Initially, we collected all eligible studies; however, the sample size of these studies is not yet large enough, especially in certain ethnic groups. Therefore, not only is the likelihood of I/II type errors high, but there is insufficient statistical capacity to assess the correlations between the 7 SNPs and PTB risk. Second, serum MBL concentration was not assessed in our study, which would have been helpful for detecting and understanding the mechanism of SNPs in the  article  includedMBL2 may be related to PTB risk. Larger sample sizes and additional gene-environment interactions should be considered in future studies.In summary, our study indicated that the rs11003125 and A/O-combined SNPs in


Metabolic syndrome (MetS) has been related to the pathogenesis of variety categories of cancers. This meta-analysis aimed to determine the association between MetS and the incidence of lung cancer.2 statistic were used to analyze the heterogeneity. Random-effect model which incorporates the potential heterogeneity was used for the meta-analysis.Relevant cohort studies were identified by search of PubMed, Embase, and Cochrane’s Library databases. Cochrane’s Q test and I2 = 0) nor the IDF criteria  was associated with an affected risk of lung cancer. Subgroup analyses showed consistent results in women and in men, in studies performed in Asian and non-Asian countries, and in prospective and retrospective cohorts . Meta-analysis limited to studies with the adjustment of smoking status also showed similar results . No publication bias was detected based on the Egger regression test (p = 0.32).Five cohort studies with 188,970 participants were included. A total of 1,295 lung cancer cases occurred during follow-up. Meta-analyses showed that neither MetS defined by the revised NCEP-ATP III criteria (hazard ratio [HR]: 0.94, 95% confidence interval [CI]: 0.84 to 1.05, p = 0.25; ICurrent evidence from cohort studies does not support that MetS is an independent risk factor for the incidence of lung cancer. Metabolic syndrome (MetS) is a cluster of metabolic disorders characterized by the pathophysiological presence of central obesity, insulin resistance, high blood pressure, and dyslipidemia . With thLung cancer is one of the most common cancers. In 2012, there were about 1.8 million new lung cancer cases and 1.6 million cases of death, respectively, accounting for about 13% of the total number of cancer diagnosis and 20% of the total number of cancer deaths . SmokingWe performed the meta-analysis in accordance with the MOOSE   and CochDatabases of PubMed, Embase and Cochrane’s Library were searched for relevant records. As for the search strategy, the combined terms were entered into the above databases as a single search, as ("metabolic syndrome" OR "insulin resistance syndrome" OR "syndrome X") AND ("lung" OR "pulmonary" OR "respiratory") AND ("cancer" OR "neoplasm" OR "carcinoma") AND ("prospective" OR "prospectively" OR "retrospective" OR "retrospectively" OR "followed" OR "follow-up" OR "cohort" OR "cohorts" OR "risk" OR "incidence"). We used this keywords search strategy instead of those searched as "text words" or as "Mesh terms" or “Emtree” to retrieve more comprehensive records. The search was limited to human studies published in English language. The reference lists of original and review articles were also analyzed using a manual approach. The final literature search was performed on April 20, 2020.Articles were included in the meta-analysis if they met all the following criteria: (1) published as full-length article in English; (2) reported as cohort studies  with the follow-up duration of at least one year; (3) included adult population (≥ 18 years of age) without lung cancer at baseline; (4) MetS defined according to the criteria of the original articles was identified as exposure of interest at baseline; (5) participants without MetS at baseline was considered as controls; (6) documented the incidences of lung cancer during follow-up; and (7) reported the adjusted hazard ratios  and their corresponding 95% confidence intervals (CIs) for the incidence of lung cancer comparing individuals with MetS at baseline to those without MetS. Reviews, letters, editorials, preclinical studies and non-cohort studies were excluded.Two authors independently performed literature searching, data extraction, and quality assessment according to the predefined inclusion criteria. Discrepancies were resolved by consensus. Data that were extracted include: (1) name of first author, year of publication and country where the study was performed; (2) design characteristics (prospective or retrospective); (3) characteristics and numbers of the participants; (4) criteria for the diagnosis of MetS; (5) follow-up period; (6) number of lung cancer cases in each study; and (7) variables adjusted when presenting the results. The quality of each study was evaluated using the Newcastle–Ottawa Scale  which ra2 test were used to evaluate the heterogeneity among the include cohort studies [2 > 50%. We used a random-effect model to synthesize the HR data because this model is considered as a more generalized method which incorporates of the potential heterogeneity [We used HRs as the general measure for the association between MetS at baseline and the incidence of lung cancer. Data of HRs and their corresponding stand errors (SEs) were calculated from 95% CIs or p values, and were logarithmically transformed to stabilize variance and normalized the distribution . The Coc studies . A signiogeneity . Sensitiogeneity . Predefiogeneity , we evalogeneity . We usedThe processes of database searching were presented in Fig. Table The characteristics of the included cohort studies were presented in Table 2 = 0). Results of sensitivity analyses by excluding one study at a time did not significantly affect the result, suggesting the stability of the main result (Table 2 = 0). Results of subgroup analyses according to the gender of the participants were also similar . These results suggested that current evidence from cohort studies does not support that MetS is an independent risk factor for the incidence of lung cancer.Results of our meta-analysis may reflect the fact that the components of MetS may have different influences on the risk of lung cancer. For the association between obesity and lung cancer risk, current evidence is not consistent. A previous meta-analysis 31 studies showed that obesity is protective factor against lung cancer , whereasThe strengths of our study, compared with the previous meta-analysis  may inclIn conclusion, results of our meta-analysis showed that current evidence from cohort studies does not support that MetS is an independent risk factor for the incidence of lung cancer. The influences of each component of MetS on pathogenesis of lung cancer should be evaluated in future studies.


N-acetyltransferase 2 (NAT2) acetylator status can be classified into three groups depending on the number of rapid alleles : rapid, intermediate, and slow acetylators. Such acetylator status may influence the occurrence of adverse drug reactions (ADRs) during sulfasalazine treatment. This systematic review and meta-analysis aimed to evaluate the association between NAT2 acetylator status and ADRs of sulfasalazine. We searched for qualified studies in PubMed, Web of Science, Embase, and the Cochrane Library. Odds ratio (OR) and 95% confidence intervals (CIs) were calculated to evaluate the strength of the association between NAT2 acetylator status and ADRs of sulfasalazine. Nine cohort studies involving 1,077 patients were included in the meta-analysis. NAT2 slow acetylators were associated with an increase in overall ADRs , discontinuation due to overall ADRs , and dose-related ADRs , compared with rapid and intermediate acetylators. In conclusion, NAT2 slow acetylators are at risk of ADRs during sulfasalazine treatment. Based on our findings, NAT2 genotyping may be useful to predict the occurrence of ADRs during sulfasalazine treatment. It has also been widely used to treat other autoimmune diseases, such as ankylosing spondylitis, Crohn’s disease, and ulcerative colitis4. Although the mechanism of sulfasalazine action is not well established, the drug is known to have anti-inflammatory and immunomodulatory effects1.Sulfasalazine is one of the classical agents used to treat rheumatoid arthritis5. Severe or fatal ADRs such as hematologic disorders (including leukopenia), systemic hypersensitivity reactions, lupus-like syndromes, hepatotoxicity, and pulmonary complications can occur, even though the incidence is low6. Most ADRs generally occur within the first few months of starting sulfasalazine treatment, and about 20–30% of patients discontinue the drug during this period because of ADRs8. Thus, the occurrence of ADRs is an important factor influencing sulfasalazine treatment continuation.The common adverse drug reactions (ADRs) of sulfasalazine are gastrointestinal symptoms , headache, dizziness, and rash9. After oral administration, approximately 15–30% of sulfasalazine is absorbed in the small intestine11, and the rest is metabolized in the colon to 5-ASA and SP by bacterial azoreductase13. About 25% of 5-ASA is absorbed as the unchanged form. SP, however, which is highly associated with sulfasalazine ADRs, is mostly absorbed from the colon, acetylated by N-acetyltransferase 2 (NAT2) in the liver, and then eliminated renally12.Sulfasalazine consists of two components, 5-aminosalicylate (5-ASA) and sulfapyridine (SP), which are connected by an azo bondNAT2 gene, located on human chromosome 8, is highly polymorphic14. Although allele frequencies of the NAT2 gene differ widely across ethnicities15, NAT2*4 is regarded as the wild-type allele16. While NAT2*4, which is considered a rapid allele, maintains NAT2 activity, several mutated alleles that reduce enzyme activity have been identified ; these alleles are considered slow alleles17. As NAT2 shows excellent genotype–phenotype correlations, NAT2 genotypes can be classified into three phenotypes: rapid acetylators , intermediate acetylators  and slow acetylators 18. Or, the genotypes can be categorized into two groups depending on whether at least one NAT2*4 allele is present or not.The NAT2 acetylator status and sulfasalazine ADRs. However, study results were inconsistent, potentially because of different ethnicities and disease populations. Also, the individual studies had limited statistical power because of relatively small sample sizes. Therefore, we performed a systematic review and meta-analysis to determine the association between NAT2 acetylator status and sulfasalazine ADRs.Since NAT2 plays an important role in sulfasalazine metabolism, several previous studies assessed the association between 20, two case series22, and one case-control study23. Finally, nine studies32 with 1,077 patients were included for meta-analysis.The study selection process is shown in Fig. 33. Among five studies in which the Hardy-Weinberg equilibrium (HWE) test was performed32, two studies were not in HWE32. Quality scores evaluated by the Newcastle-Ottawa Quality Assessment Scale (NOS) ranged from 4 to 8.The main characteristics of each study are summarized in Table NAT2 SAs and RAs + IAs are shown in Fig. p = 0.005). Since significant heterogeneity was observed , a random-effects model was used. For secondary outcomes, SAs had significantly increased risks of discontinuation due to overall ADRs  and dose-related ADRs  compared with RAs + IAs, using a fixed-effects model .The meta-analysis results comparing sulfasalazine ADRs between NAT2 acetylator statuses . However, there was no significant difference between IAs and RAs .Five studies were included in comparison of the three NAT2 SAs were compared with RAs + IAs, a significant association was found between NAT2 acetylator status and overall ADRs in Asians . However, there was no significant difference in other populations, possibly due to the small number of studies included.Subgroup analysis by ethnicity was performed for the primary outcome Fig. . When NAp = 0.0002). In the analysis using studies on rheumatoid arthritis, OR of SAs compared to RAs was 3.14 . The subgroup analysis results were similar to that from the entire meta-analysis .We additionally performed subgroup analyses on studies with prospective design and rheumatoid arthritis to rule out the confounding effects due to the different study designs, diseases, and doses. Results from studies with prospective design showed that the OR of SAs compared to RAs was 2.97 . When the Ricart et al. study25 was removed, heterogeneity was greatly reduced . Moreover, Galbraith plot showed that the studies of Ricart et al.25 and Tanaka et al.26 were the major source of heterogeneity , and heterogeneity was not observed . A funnel plot for the primary outcome is shown in Fig. p = 0.6207; Egger’s test: p = 0.392).Sensitivity analysis was conducted by sequentially excluding each study to assess the effects of individual studies on the overall meta-analysis estimate 37. According to these studies, NAT2 SAs were associated with an increased risk of AT-DILI. Moreover, it has been reported that slow acetylation is a risk factor for ADRs associated with other drugs, such as hydralazine38, amifampridine phosphate39, and co-trimoxazole40. These results could be explained by reduced NAT2 activity, which led to higher concentrations of toxic substances.The present results are consistent with those of several previous meta-analyses, which assessed the association between 41. Autoimmune disease itself has diverse clinical manifestations and the etiology remains elusive43. In addition, the detailed mechanisms of adverse reactions of drugs used for autoimmune diseases have not been understood; however, SP, a sulfonamide metabolite of sulfasalazine, is known to be a main cause of ADRs during sulfasalazine treatment6. This moiety might cause sulfa-related toxicity including headache, nausea, and vomiting as well as several allergic reactions44. These ADRs seemed to be dose related and mostly reversible, being affected by hepatic acetylation5.Autoimmune diseases, including rheumatoid arthritis, show heterogeneity in both pathophysiological and clinical aspects45. As SP undergoes N-acetylation or glucuronidation, followed by hydroxylation, and is then eliminated renally, N-acetylation is the main route of metabolism13. A previous pharmacokinetic study reported that NAT2 SAs had higher serum concentrations of SP than RAs + IAs46, suggesting that ADRs in SAs are associated with higher concentrations of SP. In addition, several NAT2 phenotyping studies, which determined acetylator phenotype from the serum concentration ratio of free SP to total SP using Evan’s method47, instead of genotyping, reported that SAs experienced significantly more ADRs than RAs49. Therefore, the N-acetylation rate of SP is a possible key factor in the occurrence of ADRs.A previous study demonstrated that patients with ADRs during sulfasalazine treatment had increased serum concentrations of SP (>50 μg/mL); however, such increased concentrations were not observed for sulfasalazine, 5-ASA, and metabolites of SPNAT2 SAs were more likely to experience overall ADRs than RAs or IAs, whereas no significant difference was found between RAs and IAs. A previous meta-analysis showed similar results: Shi et al.36 found that IAs were not significantly associated with an increased risk of AT-DILI. The results suggested that having only one rapid allele may not be a risk factor for ADRs because this is adequate for metabolizing drugs and toxic substances.Additionally, in comparisons of the three acetylator statuses, 50. Although the small number of studies  could make interpretation difficult in advanced tools such as Begg’s or Egger’s test51, Herbison et al. reported that meta-analysis with as few as four or five studies could produce robust results consistent with long-term results52. Second, participants’ diseases, ADR definitions, and SNPs used for genotyping varied among individual studies, thereby increasing heterogeneity. Third, some potential risk factors, such as sulfasalazine dose, treatment duration, and concomitant medications could not be adjusted due to lack of information from individual studies. Fourth, since only studies published in English were included in the meta-analysis, the possibility of publication bias could not be excluded, even though statistical tests demonstrated that there was no evidence of publication bias.This meta-analysis may still have some limitations that should be considered when interpreting the results. First, there was a limited number of studies, especially in non-Asian populations. However, according to a typical Cochrane review, the median number of included studies per review was 6 (interquartile range: 3–12)NAT2 acetylator status and sulfasalazine ADRs. By combining inconsistent results from individual studies, we could draw the conclusion that NAT2 SAs have an increased risk of ADRs during sulfasalazine treatment, especially in Asian populations. Therefore, to prevent the occurrence of ADRs, individualized sulfasalazine treatment strategies according to NAT2 genotyping can be proposed based on our findings.To our knowledge, this is the first systematic review and meta-analysis to evaluate the association between NAT2 acetylator status and sulfasalazine ADRs. The following search terms were used:  AND (N-acetyltransferase 2 OR N-acetyl-transferase 2 OR N acetyltransferase 2 OR Nacetyltransferase 2 OR NAT2 OR arylamine acetyltransferase) AND (polymorph* OR variant* OR mutation* OR genotyp* OR phenotyp* OR haplotyp*). The search was not restricted by publication date. Duplicates and irrelevant studies were removed through the initial screening of titles and abstracts according to the eligibility criteria.Two researchers independently searched four databases  on 11 July 2019, for studies about the association between NAT2 acetylator status and ADRs in sulfasalazine-treated patients; (2) using prospective or retrospective cohort study design; (3) providing sufficient information to calculate OR and 95% CIs; and (4) being published in English. Exclusion criteria comprised: (1) conference or meeting abstracts, summaries, reviews, comments, letters, news, and editorials; (2) in vitro or animal studies; or (3) studies in healthy volunteers. If there were overlapping data, only the most recent and comprehensive data were included in the meta-analysis.The following criteria were used to identify eligible studies: (1) evaluating the association between et al.5, and these reactions included nausea, vomiting, headache, malaise, hemolytic anemia, reticulocytosis, and methemoglobinemia.All data were extracted independently by two researchers, and discrepancies were resolved by consensus. The following information was extracted from each study: name of the first author, publication year, ethnicity, study design, disease of patients, sulfasalazine dose, genotyping method, single nucleotide polymorphisms (SNPs) used for genotyping, and deviation from HWE. Also, the following outcome data were extracted from each study: the number of patients with or without overall ADRs (primary outcome), the number of patients who discontinued the drug due to overall ADRs, and the number of patients who experienced dose-related ADRs (secondary outcomes). Dose-related ADRs of sulfasalazine were defined based on a study by Taffet 53. There are three categories in NOS: selection of study sample, comparability between case and control group, and outcome assessment. Each study can be assessed with a total score of 0–9. In this review, we rated one point in each item of comparability, if age and other known risk factors  were matched or adjusted in the analysis. The minimum follow-up period was 12 weeks, which was considered sufficient for outcomes to occur8.Two researchers independently assessed the selected studies based on the NOS for cohort studiesNAT2 acetylator status and sulfasalazine ADRs. NAT2 SAs  were compared with RAs + IAs for each outcome. Additionally, three comparisons were performed for overall ADRs: SAs vs RAs, SAs vs IAs, and IAs vs RAs. A p-value <0.05 was considered statistically significant. Heterogeneity between studies was assessed by a chi square-based Q test and an I2 test; I2 > 50% was considered to indicate significant heterogeneity. When there was no statistical evidence of heterogeneity, the fixed-effects model (Mantel-Haenszel method) was used, otherwise the random-effects model (DerSimonian-Laird method) was used to calculate pooled estimates55. If a study had no events in both comparison groups, the study was excluded from meta-analysis of that outcome. Subgroup analysis was performed according to ethnicity.OR and 95% CIs were calculated by Z test to estimate the strength of the association between 56. Publication bias was assessed using funnel plots, Begg’s test, and Egger’s test. When a p-value was <0.05, we considered it statistically significant for publication bias. All statistical analyses were performed using Review Manager  and R software . The review followed Preferred Reporting Items for Systematic Reviews and Meta-analyses (PRISMA) guidelines57.To assess the stability of the results, sensitivity analysis was performed by sequentially excluding each study, or by omitting outlier studies. Galbraith plot was used to spot outliers as potential sources of heterogeneitySupplementary Information.