Many studies have showed that diabetes mellitus (DM) might be a risk factor for certain types of cancers. However, there are still inconsistent results on the effects of DM on the risk of esophageal cancer (EC). The objective of this study is to investigate the association and to quantify the correlation between DM and EC by a meta-analysis.I2 value was used to assess heterogeneity, and funnel plot analysis was for publication bias.The initial search identified 339 articles. Those publications that did not report the exact number of EC cases were removed. Finally, 13 meaningful studies were extracted from the databases of PubMed, MEDLINE, and Web of Science. All pooled analyses of risk ratios (RRs) and 95% confidence intervals (CIs) were assessed by a random-effect or fixed-effect model. Subgroup analysis was implemented on the basis of the sex or ethnicity. P < .001). Subgroup analysis based on gender showed that male was an important risk factor for EC , but female was not . In addition, subgroup analysis based on ethnicity showed that DM was significantly correlated to EC in North America subjects , and in Europe subjects , whereas no correlation was found in Asian subjects . Furthermore, DM had a correlation to an increased risk of esophageal adenocarcinoma (EAC) .The result showed that there was a positive correlation between type 2 diabetes mellitus (T2DM) and EC risk (RR = 1.28, 95% CI: 1.12–1.47, This meta-analysis indicates that DM is positively correlated to EC. However, the results should be interpreted with caution because of the limitations on potential clinical confounding factors in each study included in this meta-analysis. Generally, EC has 2 types: esophageal squamous cell carcinoma (ESCC) and esophageal adenocarcinoma (EAC). In contrast with ESCC, EAC was the more common type. In the 1990s, more than 95% of esophageal malignancies were ESCC. However, the incidence of EAC increased approximately 6-fold in the United States from 1975 to 2001 and a decreased incidence of ESCC has been seen in western countries. In addition, the survival rate of EC remains low, despite the rapid progress of diagnosis and treatment. Therefore, EC remains a seriously fatal disease around the world.EC has been a serious disease around the world. It is said that the rate of 5-year mortality will exceed 80% in patients with EC. that affect the progression of EC would be important for prevention or early detection of EC. EC has 2 main types, and there are several significant differences in pathogenesis, tumor biology, and individual characteristics between ESCC and EAC. For ESCC, tobacco, dietary carcinogen exposure, and alcohol consumption are the known risk factors. Besides, Stroup et al also identified the infection of Helicobacter pylori as another potential risk factor. For EAC, gastro-esophageal reflux disease (GERD), white race, male gender, obesity, and smoking are regarded as established risk factors for EAC development. Nevertheless, many patients have not been exposed to these risk factors.Prevention of EC seems to be very urgent. And identification of potential risk factors,11 Many studies have demonstrated that DM might be a risk factor for some types of cancer, such as hepatocellular carcinoma, breast cancer, endometrial cancer, bladder cancer, and kidney cancer.–16 The detailed mechanisms may be explained by the effects of insulin and insulin-like growth factors (IGFs) axis on cells growth. A number of studies reported that IGFs axis could trigger intracellular signaling transduction involved in the development of cancers.,18In recent years, the prevalence of diabetes mellitus (DM) was significantly increased around the world.,16,19–29 However, results of different studies are inconsistent. This study aims to resolve the inconsistencies by incorporating some of the relevant studies investigating the exact number of EC patients in DM and non-DM groups.In 1991, the association between type 2 diabetes mellitus (T2DM) and EC was first reported. After that, a number of studies regarding DM and EC susceptibility were conducted.22.1 All articles involving DM and EC were obtained from the English literature. A computerized literature search was implemented in PubMed, MEDLINE, and Web of Science databases till August 2016 with the following text word or Medical Subject Heading (MeSH) terms: (“diabetes mellitus” or “diabetes” or “DM” or “T2DM”) AND .This meta-analysis was conducted according to Preferred Reporting Items for Systematic Reviews and Meta-analyses guidelines (PRISMA).2.2The studies that met all the following criteria were brought into the meta-analysis: they were case–control studies; studies must have incidence of EC, EAC, or ESCC ; there were odds ratio and 95% confidence interval (CI) in case–control studies; and patients diagnosed with type 1 diabetes mellitus (T1DM) were excluded from this analysis.2.3As our study was a secondary analysis regarding human subject data published in the public domain, ethical approval was not necessary in this review.2.4Three investigators  independently collected the following data from the included studies according to a predefined protocol: the first author's last name, publication time, country of the study, source, ethnicity , sample size (cases and controls), confounders or adjusted factors, the odds ratio and 95% CI, and solved disagreement by discussion. The NOS system consisted of 3 aspects: subject selection (0–4 points), comparability (0–2 points), and clinical outcome (0–3 points). And NOS scores ≥6 are identified as high-quality studies.Three investigators  independently accomplished the quality assessment according to the Newcastle–Ottawa Quality Assessment Scale (NOS).2.5The quality of evidence based on the results of meta-analysis would be assessed by GRADE system (version 3.6). The GRADE system included 4 levels of evidence: high quality: further research may not change the effect of the credibility of the evaluation result; the medium quality: further research is likely to affect the curative effect of the credibility of the evaluation result and can change the results of the assessment; low quality: further studies are likely to influence the reliability of the efficacy evaluation, and this assessment is likely to change; and extremely low quality: any efficacy evaluation results are very uncertain. Two reviewers (BX and CZY) assessed quality independently and solved disagreement by discussion.2.6 using Cochrane Library software Review Manager (version 5.3 software). I2 value was adopted for the quantification of statistical inconsistency between studies due to heterogeneity, in which I2 < 30% demonstrated mild heterogeneity, I2 is between 30% and 70% showed moderate heterogeneity, and I2 > 70% suggested severe heterogeneity. Funnel plot analysis was used to assess publication bias using Stata software (version 12.0). P < .05 was considered statistically significant.The incidence of EC in each study was regarded as a binary variable. Study-specific risk ratio (RR) was assessed by a fixed- or random-effect model3,16,19–29 were included in this meta-analysis , we chose the random-effect model to estimate the pooled RR. The results showed a meaningful association between DM and EC, suggesting DM was a risk factor for EC  , but female was not  (Table P < .001) and in European subjects ; however, no correlation was found in Asian subjects   (Table P < .001) when controlled for body mass index with a meta-analysis including 7 studies. Similarly, a significant correlation was also found between diabetes and EC risk, when we limited the meta-analysis to the studies that controlled for smoking habits  or alcohol use .When we limited the meta-analysis to the 6 studies specified for the subtypes of EAC,P = .951) and the Egger test (P = .742) indicated that there was no publication bias in the present study  Hyperglycemia can cause interaction of the crosstalk between oxidative stress and advanced glycation products (AGEs)/advanced glycation products receptor (RAGE) system. The crosstalk can activate cell signaling pathways that stimulate cellular growth and inhibit apoptosis, which could lead to cancer and cell invasion.Hyperglycemia is the connecting link between DM and cancer.(2) What is more, the study demonstrated that serum total and free IGF-1 levels are significantly increased among viscerally obese patients compared with normal weight controls, and among patients with EAC compared with Barrett esophagus (BE) and healthy controls.Type 2 diabetes has a correlation with hyperinsulinemia and insulin resistance. Hyperinsulinemia is associated with increased bioactive serum IGF-1 and can upregulate the concentration of IGF-1. And activation of the IGF-1 and insulin receptors results in intracellular signaling cascades in both the extracellular signal-related kinase (ERK) and phosphatidylinositol-3 kinase (PI3K) pathways, producing downstream mitogenic, antiapoptotic, and proangiogenic effects that may favor tumor growth in EAC.(3)Delayed gastric emptying commonly seen in DM may be another mechanism.(4)Gastric hypomotility is a key factor for reflux symptom development and progression which may lead to BE and EAC.(5),14 In a word, each of these mechanisms proves a metabolic abnormality in DM patients and could underlie the association between EC and DM.Obesity which is common in T2DM is an established risk factor for EAC, although the precise mechanism remains unknown. An emerging hypothesis is that metabolic aberrations accompanied with obesity lead to changes in hormones and cytokines, including insulin, and serum insulin may play a role in BE progression through a number of mechanisms.As we all know, DM is a very common disease, and its prevalence has been significantly increased around the world. Although there is no study investigating this pathway within a single population so far, a meta-analysis by Cook et al has proved that BE, a potentially precancerous condition, has a pooled male/female sex ratio of 1.96/1 ; and EAC has a higher sex ratio (7–10:1) in most western countries. This difference of sex ratio might be in part explained by the gonadal hormones , because previous studies demonstrated that compared to normal cells, the expression of sex hormone receptors has been upregulated in human EAC cells.,46 In a word, more researches on the effects of gender on EC with DM needed to be done for underlying mechanisms and therapeutic interventions. In addition, we also found the North American and European subjects had an increased risk of EC compared with Asian subjects. This may partly result from genetic diversity among ethnicities. Furthermore, lifestyle  could inevitably play an important role in this process. What is more, the prevalence of obesity in western countries is higher than that in eastern countries, which is a risk factor of GERD that could lead to EAC.,14In the present meta-analysis, we found that male with DM had an increased risk of EC. But the detailed mechanisms involved in this phenomenon are still unclear. As we all know, male had a predominant position among patients with GERD, BE, and EC, especially EAC. Each of these may enhance difficulty to assess EC risk in DM patients, so deep-going researches based on these confounding factors are needed.There are still several limitations in present meta-analysis. First, case–control studies are susceptible to recall and interviewer bias. And the assessment of exposure factors in the studies was based on “self-statement,” which would introduce bias into these studies and affect the real results. Second, the articles in the present meta-analysis did not distinguish T1DM from T2DM. Since the misclassification of diabetes exists, it will weaken the degree of the correlation between diabetes and EC risk. However, it is likely that the majority of DM patients in this meta-analysis are T2DM, as this disease is by far the most common form particularly in older persons and the incidence of T1DM is less than T2DM. Third, confounding factor is likely to be present. Unhealthy lifestyles  and obesity have been considered to increase the risk of cancers. Although most studies in this meta-analysis have controlled these lifestyle factors, not all interference factors could be completely ruled out. Moreover, no researches could exclude the effects of hypoglycemic drugs, which may increase or decrease cancer risks.,27,34,35 which showed a significant positive correlation between diabetes and EAC risk. However, the results need to be interpreted with caution because of several reasons: The incidence of diabetes among controls was different from prevailing population estimates for each study; Five studies adjusted for GERD, which are the most important risk for EAC. And several studies did not control for smoking habits and body mass index, which are also risk factors for EAC. The association may reflect confounding by the 2 risk factors. Therefore, further researches are to be conducted on the correlation between DM and EAC.In this meta-analysis, there were 6 studies presenting results specific for EAC,According to the methodology of GRADE quality evaluation, the quality of the evidence was very low in overall study result and most subgroup analysis results, and there were several possible reasons: the weak study design was unlikely to explain all of the apparent benefit or harm, even though observational studies were likely to provide an overestimate of the true effect; the assessment of exposure factors in the studies was based on “self-statement,” which would introduce bias into these studies and affect the real results. Therefore, comprehensive therapeutic interventions should be seen in DM patients. As the association between DM and EC remains in dispute, a well-designed prospective study on the correlation between DM and EC is still to be warranted.In conclusion, this meta-analysis indicates that DM may have an association with an increased EC risk. However, it is a pity that most of DM patients only focus on DM treatment other than prevention of cancers or screening for several types of cancers.


A linear and curvilinear trend was observed in dose–response meta-analysis between CRA risk in association with PhIP, MDM, and MeIQx. CRC risk  was increased by uptake of MeIQx , DiMeIQx  and MDM . No publication bias could be detected, whereas heterogeneity was in some cases rather high. Mutagenic compounds formed during cooking of meat at high temperature may be responsible of its carcinogenicity. Much evidence suggests that the positive association between meat intake and colorectal adenoma (CRA) and cancer (CRC) risk is mediated by mutagenic compounds generated during cooking at high temperature. A number of epidemiological studies have estimated the effect of meat-related mutagens intake on CRC/CRA risk with contradictory and sometimes inconsistent results. A literature search was carried out  to identify articles reporting the relationship between the intake of meat-related mutagens , 2-amino-3,8-dimethylimidazo quinoxaline (MeIQx), 2-amino-3,4,8-trimethylimidazo quinoxaline: DiMeIQx, benzo(a) pyrene (B(a)P) and “meat derived mutagenic activity” (MDM)) and CRC/CRA risk. A random-effect model was used to calculate the risk association. Thirty-nine studies were included in the systematic review and meta-analysis. Polled CRA risk (15229 cases) was significantly increased by intake of PhIP (OR = 1.20; 95% CI: 1.13,1.28;  Colorectal cancer (CRC) is the second most common cancer in women and third in men worldwide . In 2012Various mechanisms have been proposed to explain the effect of meat intake on CRC risk . Meat isN-acetyltransferases, while glucuronidation by UDP-glucuronosyltransferases (UGT) is the primary route of HCAs detoxification [HCAs are formed from the reaction, at high temperatures, between creatine or creatinine (found in muscle meats), amino acids, and sugars. Their formation depends upon the type of meat, the cooking method, the temperature and the duration of cooking ,10. HCAsfication ,10. Becafication . More refication . Initialfication . Subsequfication ,17,18,19We conducted, therefore, a systematic review of the literature to shed light on the relation between HCAs intake and CRC/CRA risk. The investigation was extended to both the “meat derived mutagenic activity” (MDM) and the B(a)P intake because in several studies on HCAs the association of CRC/CRA risk with these two parameters was also reported. In addition, we carried out for the first time, a meta-analysis to provide quantitative estimates of the association.In this study, the standard procedures for conducting and reporting meta-analysis according to MOOSE  guidelines and PRISMA  statement were followed ,21. A comprehensive literature search, without restrictions, until 27 January 2017 through PubMed and Web of Science was carried out to identify all the original articles on the association between HCAs intake and colorectal cancer and adenoma risk. The following search key words were used: (“heterocyclic amines” OR PhIP OR MeIQx OR DiMeIQx OR “mutagen index” OR “benzo(a)pyrene”) AND (neoplasm OR cancer OR “neoplastic disease” OR neoplasia OR tumor OR adenoma) AND . Similarly, with the same key words, the Scopus database was also accessed to find eventually missing items. Furthermore, the reference lists of included articles and recent relevant reviews were manually examined to identify additional relevant publications.We focused on studies in which the amount of HCAs  and B(a)P consumed by different groups was clearly reported. Potential identified articles were included if they met the following criteria: (i) used a case-control or prospective study design; (ii) evaluated the association between HCAs intake and colon, rectal and colorectal cancer (CRC) and adenomas (CRA) risk; and (iii) presented odds ratio (OR), relative risk (RR) or hazard ratio (HR) estimates with 95% confidence intervals (CIs). When there were several publications from the same study, the publication with the largest number of cases was selected. Although useful to have background information, reviews and meta-analysis were excluded. No studies were excluded for weakness of design or data quality.After removal of duplicates, two independent investigators carried out the selection process by screening titles and abstracts for all potentially eligible studies. Disagreements between evaluators were resolved by discussion or in consultation with a third author. From the selected studies we extracted the following data: the first author’s last name, year of publication, study region and design, tumor site, sample size , age, duration of follow-up for cohort studies, dietary assessment method and HCAs quantification method, HCAs doses comparisons, OR/RR/HR estimates with 95% confidence intervals for the highest versus lowest category of HCAs intake, matched or adjusted variables. When multiple estimates were reported in the article, we abstracted those that adjusted for the most confounding factors. The study quality was assessed by a 9-star system based on the Newcastle–Ottawa Scale method (NOS) . The fulp < 0.05 was considered statistically significant. The overall effect-size statistic estimated was the average of the logarithm of the observed odds ratio (approximated to RR when necessary) associated to the highest versus the lowest level of HCAs consumption. We used the results of the original studies from multivariable models with the most complete adjustment for potential confounders. We used random effects model and the inverse variance weighted method to calculate summary OR and 95% confidence intervals. A two-tailed p-value for curve linearity or non-linearity was calculated by testing the null hypothesis that the coefficient of the second spline is equal to zero. Linear trends with fixed effects were also tested. We further aimed to quantify the effects of HCAs consumption on risk of CRA and CRC by a dose–response meta-analysis with random-effect models ,24. The Q test and used the I2 statistic to quantify the proportion of the total variation due to that heterogeneity [Q statistic, a p-value < 0.05 was considered to be representative of statistically significant heterogeneity. The I2 statistic yields results ranged from 0 to 100%  [I2 was greater than 75% considering these adjustments we decided to consider just the original published results. We evaluated heterogeneity between studies with Cochran’s ogeneity . For thegeneity) . To explp-value of less than 0.05. Subgroup analysis were conducted for the case-control and cohort studies.Results of the meta-analysis may be biased if the probability of a study being published is dependent on its results. We used the methods of Begg and Mazumdar, and Egger et al. to detect publication bias . Both mep ≤ 0.05 were considered significant.The ProMeta Version 2.0 statistical program  and packages dosresmeta 1.3.2. for R 3.1.2.  was used for the analysis . All repn = 714) and Web of Science (n = 1066) databases and after removing duplicate (n = 508), 1272 records were identified for title and abstract revision  considering the association of meat mutagens with CRA are shown in n = 3) the total HCAs uptake was not significantly associated with CRA risk P, MDM and total HCAs are shown in CRA risk F. I2 = 42.37%, p = 0.022) (p = 0.002)  and MDM . Stratification of the polled data based on anatomic site showed that the highest effects were observed on colorectal adenoma. Due to the low number of data, the risk of rectal adenoma was calculated only for the exposure to PhIP. An increment of risk was observed but it was not statistically significant  . When st= 0.002) . Heterognificant . Howevernificant .p = 0.073) when was removed the study of Sinha et al., 2005 [No evidence of publication bias could be detected for risk in any case, as evidenced by both the Egger and Begg tests , and funl., 2005 .p < 0.0001) for 2000 revertants/day and 1.42  for 7000 revertants/day of MDM intake (I2 = 9%) (p = 0.035) for 50 ng/day of MeIQx intake (I2 = 25.8%) (p = 0.04) for 100 ng/day increment of PhIP intake (I2 = 70%). There were no other statistically significant dose–response results  A. The do= 25.8%) B. Furthe= 25.8%) C. The esGeneral characteristics of the 22 selected studies (19 case-control and 3 cohort) considering the association of meat mutagens with CRC are shown in I2 = 60.66%, p = 0.0001) and DiMeIQx  but not for MDM   and MDM with no heterogeneity  P, MDM and total HCAs, and the CRC risk are shown in = 0.902) . Further= 0.902) . On the = 0.902) . In addi= 0.527) . No evidence of publication bias could be detected for risk in any case as evidenced by both the Egger and Begg tests  and funnFor the dose–response analysis eight studies were considered ,65,67,68Many epidemiological studies have investigated the effect of meat consumption on colorectal cancer risk. The results of these studies have been revised and analyzed in several meta-analysis which have concluded that the intake of red or processed meat is associated to an increased CRA and CRC risk ,3. HowevThe analysis of data stratified based on neoplastic site  also showed several discrepancies between the effects exerted by the different mutagens on CRA and CRC. For instance, while PhIP, MeIQx, DiMeIQx and B(a)P significantly increased the adenoma risk in colorectal site, no effects could be observed toward carcinoma risk. These results could seem unexpected because experimental and observational studies indicate that adenoma is a valid surrogate end point for carcinoma . AccordiRegarding the CRC risk in different anatomical subsites, we observed that in the colon the risk was increased by MeIQx, DiMeIQx and MDM while the same compounds did not affect the rectal site. This observation indicates that either the mutagens may be acting at different location within the colorectum or that various anatomical subsites have a different sensitivity toward the mutagens. Indeed, much evidence suggests that the CRC risk associated to various environmental and genetic factors, as well as the sensitivities toward chemotherapeutic drugs, is different for proximal and distal tumors . InteresThe importance of the whole diet in the assessment of cancer risk is supported by several recent systematic review and meta-analysis which have considered the effects of dietary patterns instead of single components on CRC risk ,81,82. HIn conclusion, this is the first meta-analysis showing that meat related mutagens may act as a risk factors of colorectal neoplasia. Our results suggest that a possible molecular mechanisms by which meat increase the colorectal cancer risk are mediated by the presence of HCAs and PAHs that are formed during cooking at high temperature. Further studies are necessary to support these findings on different populations considering more appropriately both the dose–response relationship and the effects of potential confounders such as smoking.


Several meta-analyses have indicated that periodontal disease (PD) are related to cardiovascular diseases (CVDs). However, the association between PD and myocardial infarction (MI) remains controversial. Here we aimed to assess the association between PD and MI by meta-analysis of observational studies.PubMed, EMBASE and the Cochrane Library were searched through July, 2016. Observational studies including cohort, cross-sectional and case–control studies reporting odds ratio (OR) or relative risk (RR) with 95% confidence intervals (CIs) were included in the analysis. Either fixed or random-effects model were applied to evaluate the pooled risk estimates. Sensitivity and subgroup analyses were also carried out to identify the sources of heterogeneity. Publication bias was assessed by the Begg’s, Egger’s test and funnel plot.We included 22 observational studies with 4 cohort, 6 cross-sectional and 12 case–control studies, including 129,630 participants. Patients with PD have increased risk of MI . Substantial heterogeneity in risk estimates was revealed. Subgroup analyses showed that the higher risk of MI in PD patients exists in both cross-sectional studies  and case–control studies , and marginally in cohort studies . Further, subgroup meta-analyses by location, PD exposure, participant number, and study quality showed that PD was significantly associated with elevated risk of MI.Our meta-analysis suggested that PD is associated with increased risk of future MI. However, the causative relation between PD and MI remains not established based on the pooled estimates from observational studies and more studies are warranted.The online version of this article (doi:10.1186/s12872-017-0480-y) contains supplementary material, which is available to authorized users. Periodontal disease (PD) is defined as chronic inflammation of the supporting structure of the teeth. PD is estimated to be detectable in 20-50% of the general population and a major cause of tooth loss in adults . The symA great body of evidence has investigated the association between PD and CVD. Even though a number of meta-analyses indicated the association between PD and CVD, the cause-and-effect relationship between them was still not established with current available evidence according to an American Heart Association scientific statement . MeanwhiIn 1989, Mattila et al. first reported oral health including PD was related to acute MI . Since tThe purpose of this study was to examine the association between PD and MI in observational studies. By pooling data from individual studies and using meta-analysis, the statistical estimates of the association between PD and incidence of MI were obtained.The meta-analysis was performed following the guidelines from Meta-Analysis of Observational Studies in Epidemiology (MOOSE) group  and the Studies included for the meta-analysis if they referred to observational studies including cross-sectional, case–control, cohort and nested case–control studies. The article should estimate the association between PD and MI and provide the odds ratio (OR) or relative risk (RR) with the corresponding 95% confidence intervals (CIs).Pubmed, EMBASE and the Cochrane Library Databases were searched to identify relevant studies through July 2016. The medical subject headings (MeSH terms) as well as free text terms were used in the search strategy. The terms applied were as follows:  AND   screened the titles and abstracts of the search results and articles were excluded if they did not meet the above-mentioned inclusion criteria.Newcastle-Ottawa Quality Assessment Scale was used to assess the quality of the observational studies, including case–control and cohort studies  and the To rank the cross-sectional studies, the AHRQ was quantified in the following way. An item would be scored ‘0’, if it was answered ‘NO’ or ‘UNCLEAR’. If it was ‘YES’, then this item was scored ‘1’. Article quality was ranked as follows: low quality = 0-3; moderate quality = 4-7; high quality = 8-11 .P values and adjusting factors.Two reviewers (S. Xu and X. Liu) extracted the data from all included studies independently. Any discrepancies or uncertainties between the reviewers were resolved by consensus after rechecking the source and discussion with the third reviewer (Z. Qin). A data collection form was used to compile extracted study information and included the following items: first author’s surname, publication year, country of origin, sample size, population, study design, enrolled period, age ranges, PD exposure parameter, adjusted OR/RR and their 95% CI, PD was defined to include self-report diagnosis, or any measure of disease according to clinical, radiographic and microbiological assessment . At the very beginning, all related OR/RR based on different PD assessments were extracted. The one selected for final meta-analysis was according to the frequency of each OR/RR in all included studies. The order in this study was clinical attachment loss (CAL), pockets deep and periodontal bone loss.The OR was used in these studies, while RR and hazard ratio (HR) were considered equivalent. For one study that reported stratified OR/RR in different population, we considered each subgroup analysis as an independent study. Before pooling the data, ORs were transformed into their natural logarithm to stabilize their variance, normalize the distribution and then statistically pooled .2 metric. For the Q statistic, a P-value <0.1 was considered statistically significant heterogeneity. For I2 statistic, a value of zero was considered no heterogeneity, 25%-49% low, 50%-74% moderate, and above 75% high, respectively. To explore possible reasons for heterogeneity and test the robustness of the association, subgroup analyses were performed based on study design, methodological quality, gender, locations and PD exposure assessment parameters. When obvious heterogeneity across studies exists, the random-effect model was used to calculate the pooled estimates. Otherwise, the fixed-effect model was applied.The heterogeneity among studies was evaluated by Cochrane Q test and quantified as IPublication bias was assessed by both Begg’s rank correlation test and Egger’s linear regression test. Funnel plots were also generated to assess the publication bias.P values were 2-sided and P < 0.05 was considered statistically significantly otherwise indicated.STATA statistical software, version 11.0  was used for analysis. Initially, a total of 2558 articles were identified from PubMed 2041 articles), EMBASE (415 articles) and the Cochrane Library Database (102 articles) search. Of these citations, after evaluating the titles and abstracts, 373 studies were chosen for further review by reading the full text. Among them, 353 literatures were excluded because 286 articles were irrelevant, 14 had no OR/RR values, 28 had no clearly MI endpoint, 3 were not treated PD as exposure and 22 had no original data. Finally 22 studies from 20 articles representing 129,630 participants were included in our meta-analysis , 31. See articlesThe characteristics of each study were demonstrated in Table P < 0.001, I2 = 86.3%). Sensitivity analysis showed two studies contributed most to the heterogeneity [2 decreased to 70.68, and 73.1%, respectively. In the subgroup analysis by type of study  [2 = 16.7%, n = 6) [2 = 90.2%, n = 12) [The OR/RR from 22 studies were pooled and analyzed as presented in Fig. ogeneity . Althoug, n = 4) , 21–23, , n = 6) –20, 31 a n = 12) , 24.Fig.By removing each study at a time, sensitivity analysis was conducted to determine the influence of each study on the pooled OR. By omitting one study in turn, the pooled OR and 95% CI was not modified significantly Fig. . This seP = 0.013), Begg’s Test (P = 0.195), and Funnel plot (Fig. Additionally, publication bias was observed from Egger’s Test ( P = 0.19, and FunTo examine the stability of the studies, subgroup analyses were carried out based on several factors. Patients with PD had increased risk of MI across most subgroups by location, PD exposure parameters and study quality Table . In the In this study, we identified a statistical positive correlation between PD and MI. Patients with PD have approximately a 1-fold increase in risk for MI. Given the high prevalence of PD in population , this inSeveral meta-analyses have reported that subjects with PD have higher odds and higher risks of developing cardiovascular diseases –38. TeeuMyocardial infarction is one of the most life-threating cardiovascular diseases. Some investigations mentioned the relationship between PD and MI, while dispute exists. In 1996, Joshipural KJ’s group followed up a cohort including 44119 males for 6 years and found no overall association between PD and fatal or non-fatal MI in this population . This neIn this study, we have showed an overall increased risk for MI in PD patients. The subgroup meta-analyses by study design showed that PD was associated with MI in both cross-sectional and case–control studies, while the relation in the cohort studies was marginal. Likewise, the subgroup analyses by country, PD exposure, participant number and study quality consistently showed an apparent increased risk of MI in PD subjects. Moreover, a marginal but non-significant increase of risk was present in male population. These non-significant findings exist in cohort studies and male participants might be attributed to the small number of studies included in the subgroup meta-analyses. Intriguingly, Rydén L group in their case–control study specified that the risk of a first MI was significantly increased in patients with PD.  However,Although not fully illustrated, several mechanisms were speculated to be involved in such a relation between PD and MI. The periodontal pathogen burden and its by-products might not only contribute to the development of atherosclerosis but also the rupture of atherosclerotic plaques and occurrence of MI. Dozens of literature has demonstrated that periodontal bacteria are detected in from atherosclerotic plaques. Ohki T et al. further suggested that three species of periodontal bacteria were present in the thrombi of patients with acute MI . This fiHeterogeneity among studies was revealed in this analysis, although the relation between PD and MI has similar tendency in the overall and subgroup analysis. The significant heterogeneity was also reported in meta-analysis on PD and cardiovascular diseases , 46. By Most of the studies that exclude for the meta-analysis in this study also supported a positive association between PD and MI. A population-based study showed periodontitis was associated with an increased risk of major adverse cardiac events including MI occurrence . Other sSeveral limitation of this meta-analysis should be noted. First, limited by the nature of observational studies, the causality between PD and MI cannot be inferred from our findings. Second, as aforementioned, definitions for PD in each study included in the analysis were not identical. Although we set rules when extracting data, this discrepancy in the assessments for PD seemed not to be eliminated nor even decreased obviously. Additionally, the differences in the study design, population and criteria to measure the exposure may weaken the evidence and lead to considerable heterogeneity. Although further analysis were performed, the sources of the heterogeneity remained not clear. The meta-regression analysis might further explain the source of heterogeneity. For the paucity of the data, these analyses were not conducted. Finally, the concrete mechanisms to explain the PD and MI relation still need further investigation, though possible mechanisms were discussed above. Further prospective studies are warranted to establish the association between PD and MI, while it might also be helpful to confirm the relation with randomized, controlled interventional trials including medication to control the pain and inflammation and surgery to remove the inflamed tissue and reduce the bone damage.In conclusion, our meta-analysis yielded a statistically significant association between PD and MI. Subgroup analyses also confirmed the elevated risk for MI in PD subjects, although heterogeneity should be noted. More studies including large-scale prospective cohort studies and randomized controlled trials are warranted.


BDNF Val66Met transgenic mice have been generated, which may provide further insight into the functional impact of this polymorphism in the brain. Considering the important role of BDNF in brain function, more than 1,100 genetic studies have investigated this polymorphism in the past 15 years. Although these studies have reported some encouraging positive findings initially, most of the findings cannot be replicated in following studies. These inconsistencies in BDNF Val66Met genetic studies may be attributed to many factors such as age, sex, environmental factors, ethnicity, genetic model used for analysis, and gene–gene interaction, which are discussed in this review. We also discuss the results of recent studies that have reported the novel functions of this polymorphism. Because many BDNF polymorphisms and non-genetic factors have been implicated in the complex traits of neuropsychiatric diseases, the conventional genetic association-based method is limited to address these complex interactions. Future studies should apply data mining and machine learning techniques to determine the genetic role of BDNF in neuropsychiatric diseases.Neurotrophins have been implicated in the pathophysiology of many neuropsychiatric diseases. Brain-derived neurotrophic factor (BDNF) is the most abundant and widely distributed neurotrophin in the brain. Its Val66Met polymorphism (refSNP Cluster Report: rs6265) is a common and functional single-nucleotide polymorphism (SNP) affecting the activity-dependent release of BDNF.  Brain-derived neurotrophic factor (BDNF), a major member of the neurotrophin family, is widely expressed in the mammalian brain . The higBDNF gene is located on chromosome 11p13 and has 11 exons and 9 functional promoters that are brain region- and tissue-specific . A search with the keywords “ OR rs6265 OR (bdnf g196a) OR (bdnf 196g/a) OR (bdnf 196a/g) OR (bdnf 196 a/g)” performed in the PubMed database up to February 14, 2018 found 1,176 reports on this polymorphism . Although many reports have demonstrated the possible genetic effects of this BDNF polymorphism in diseases or brain function, other reports have failed to replicate the findings. The inconsistent findings of BDNF Val66Met genetic studies may result from many factors such as age, sex, environmental factors, ethnicity, genetic model used for analysis, and gene–gene interaction. In this review, we discuss these issues in genetic studies of the BDNF Val66Met polymorphism. We also discuss some findings for the novel function of this polymorphism.The human specific . In thisspecific . The fir in 2002 . Considediseases  . For example, converging evidence suggests that BDNF is implicated in the pathogenesis of bipolar disorder. In 2003, two research groups reported a significant association between the BDNF Val66Met polymorphism and bipolar disorder  . Therefosubjects . HoweverBDNF Val66Met polymorphism on suicidal behaviors, a meta-analysis evaluated 23 studies, including 4,532 patients and 5,364 controls, but found no evidence of an association between this polymorphism and suicidal behaviors , codominant , or recessive . The BDNF Met allelic frequency is often reported to be high in Asian populations but low in Caucasian, Central and South American, and African populations , such as the ulations . Many stpatients . This ispatients . This obpatients .BDNF Val66Met polymorphism and panic disorder is inconclusive given the mixed findings  mRNA-containing neurons during forebrain development . Evidencchanisms . First o neurons .Within the hippocampus, estrogen and BDNF both interact with a number of common receptors, enzymes and proteins such as MAP kinase, ERKs, PI3 kinase, CaMKII, CREB, and Src/Fyn . The intBDNF expression is less studied. Study in mice demonstrated that gonadectomy induced a significant decrease in the BDNF levels in the hippocampal CA1 area, which were prevented by replacement of testosterone, the major component of androgens  genetic studies. The first genetic association study of the BDNF Val66Met polymorphism and AD demonstrated that Val is the risk allele for AD , but not in men. This finding suggests that the BDNF Val66Met polymorphism has a sexually dimorphic effect on susceptibility to AD. This result is consistent with the finding that the BDNF Val66Met polymorphism has a sex-specific role  in cognitive function during normal cognitive aging  expression of BDNF and its receptors in the hippocampus and hypothalamus throughout the life span of rats have found that receptors, rather than BDNF itself, are impaired with aging . These fBDNF Val66Met polymorphism on disease susceptibility. In our studies of the BDNF Val66Met polymorphism and major depression, we found that Met carriers have an increased risk of geriatric depression, but not non-geriatric depression . The trkB signaling pathway involves many proteins that also possibly affect BDNF function. In addition, the proteolytic cleavage of proBDNF (a BDNF precursor with effects opposite to those of BDNF) to BDNF by plasmin determines the direction of BDNF action  to affect susceptibility to geriatric depression , thereby affecting AD susceptibility in women  and the BDNF Val66Met polymorphism; this interaction modulated depressive symptoms in 410 non-clinical participants , a powerful statistical tool for detecting and modeling epistasis, has been widely applied . PolygenEvidence suggests that interactions between genes and the environment influence brain development and the risk of neuropsychiatric diseases . Many enBDNF Val66Met polymorphism has been reported to interact with early life stress; thus, Val carriers with childhood trauma are more susceptible to the occurrence of subclinical psychotic experiences  demonstrated that the increased on to AD . The intBDNF DNA methylation in subjects with anxiety/depression than in controls, and this difference was more pronounced in BDNF Val66Met heterozygotes than in Val homozygotes  that recapitulates the phenotypic hallmarks of human carriers with the Met allele. BDNFMet/Met mice represent a potential model to study the biological mechanism of this polymorphism in the brain.BDNFMet/Met mice had decreased basal BDNF protein levels in the hippocampus, which could not be normalized by antidepressant (fluoxetine) administration   with a methionine . Recentlt68BDNF) . Using tt68BDNF) .BDNF Val66Met polymorphism affects the validity of these mouse models. A more precise transgenic model was developed by the Gogos group, where the mice were ‘humanized’ by inserting a small stretch of human sequence, including Val/Met at position 66  or extracellular peptidase enzymes (tPA/plasmin/MMP) into bioactive mature BDNF and pro-peptide (or pro-domain) . The ValThe BDNF pro-peptide functions as a modulator of synaptic plasticity by enhancing hippocampal long-term depression (LTD) . Mizui eBDNF Val66Met polymorphism affects the stability of the complex formed between BDNF and its pro-peptide . However, the higher activity BDNF Val allele is associated with bipolar disorder  when compared with cognition  .Furthermore, evidence suggests that increased BDNF activity has a deleterious effect and may be implicated in the pathogenesis of some diseases , 2007a,cBDNF polymorphism  might only reveal some of the BDNF genetic variability and result in the overlooking of some information from other BDNF SNPs  in the BDNF 5′ non-coding region, has been identified and reported to be associated with AD  are significantly associated with MDD.By sequencing the entire BDNF Val66Met polymorphism, this polymorphism is one of the most studied polymorphisms in neuropsychiatric diseases. However, following studies have been unable to replicate most positive findings in initial genetic studies. In this review, we highlighted critical issues in BDNF Val66Met studies, which may affect the findings of these studies. Most neuropsychiatric diseases are complex diseases that are dependent on many genetic and environmental factors that cannot be analyzed by conventional genetic association studies. Future studies should analyze various BDNF polymorphisms and these related factors by using machine learning techniques to accurately understand the genetic effect of BDNF on disease pathogenesis.Considering the important role of BDNF in the brain and the functional effect of the common The author confirms being the sole contributor of this work and approved it for publication.The author declares that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


MDM4 and gastric cancer risk in an Eastern Chinese Population. In logistic regression analysis, a significantly decreased GCa risk was associated with the rs1380576 GG variant genotype  under a recessive model, which remained significant after correction by the false-positive reporting probability. This risk was more evident in subgroups of older subjects, males, never smokers, never drinkers and cancers of non-cardia. We then performed SNP-mRNA expression correlation analysis and found that the GG variant genotype was associated with significantly decreased expression of MDM4 mRNA in normal cell lines for 44 Chinese (P=0.032 for GG vs. CC) as well as for 269 multi-ethnic subjects (P<0.0001 for GG vs. CC). Our results suggest that the MDM4 rs1380576 G variant may be markers for GCa susceptibility. Larger, independent studies are warranted to validate our findings.MDM4 is a p53-interacting protein and plays an important role in carcinogenesis. In this study of 1,077 gastric cancer (GCa) cases and 1,173 matched cancer-free controls, we investigated associations between three tagging single nucleotide polymorphisms (SNPs)  in  Helicobacter pylori (H. pylori) infection  and rs10900598 G>T in 5′-UTR) of the MDM4 gene with risk of oral cancer, squamous cell carcinoma of oropharynx and squamous cell carcinoma of the head and neck and got some positive findings nfection , and thenfection , 17. In MDM4 by using publically avaialble genotyping and mRNA expression data. The observed significantly decreased expression of the MDM4 gene associated with the rs1380576 GG variant genotype may have lead gastric cancer susceptibility in this study population. The eQTL results of the allelic effect from GTEx with stomach tissue were in the different direction from the results of lymphoblastoid cell lines, which may be due to the tissue specificity. Therefore, other studies with larger numbers of gastric tissues are needed to validate our findings. As rs1380576 is a tag SNP, this SNP is in fact in LD with several other potential functional SNPs in MDM4 that may affect the protein function at the translational level. Thus, fine mapping of this gene will be necessary to identify additional causal variants in the future.To further identify molecular mechanisms underlying our findings, we performed genotype-phenotype correlation analysis of mRNA expression levels for MDM4 tagging SNPs and gastric cancer risk with a relatively large sample size. Several limitations of our study need to be addressed. Firstly, this hospital-based case-control study may have selection bias and information bias, which may be minimized by frequency-matching cases and controls as well as the adjustment for potential confounding factors in the final analyses. Second, only three tagging SNPs of MDM4 were investigated in the present study, which did not cover all SNPs and may have missed some important variants within MDM4. Third, other risk factors, especially the H. pylori infection status, were not available for further analysis due to the nature of the retrospective study design. This limitation should be overcome in our future studies. Fourth, we were not able to measure the expression of MDM4 in mRNA and protein levels due to the lack of clinic tissues/samples. Finally, although our sample size was relatively large, the small sample size in subgroup analyses may have limited statistical power. Hence, our findings need to be confirmed by studies with larger sample sizes. Despite these limitations, our findings are biologically plausible and provide some novel clues for the role of MDM4 in the development of gastric cancer.In summary, the present study investigated the associations between three Patients with newly diagnosed, histopathologically confirmed, and untreated primary gastric adenocarcinoma were recruited between January 2009 and March 2011 at Fudan University Shanghai Cancer Center. All patients came from Eastern China, including Shanghai City, Jiangsu Province, Zhejiang Province and the surrounding regions. Participants who had gastric adenosquamous, squamous cell carcinoma, neuroendocrine tumor, stromal tumor, metastasized cancer from other organs or any histopathologic diagnosis other than gastric adenocarcinoma were excluded from this study. Cancer-free controls were recruited from a large prospective cohort recruited for Taizhou longitudinal study (TZL) at the same time period in the Eastern China with the selection criteria including no individual history of cancer. These cancer-free Han Chinese controls were frequency matched to cases on age (± 5 years) and sex . A strucGenomic DNA extraction and genotyping were conducted as described previously , with a MDM4 rs1380576 C>G SNP, we used publically available data on genotypes and transcript expression levels of MDM4 from 270 lymphoblastoid cell lines from all populations  available online for the genotype–phenotype correlation analysis [To explore the functionality of the analysis . The gent-test (for continuous variables) or Chi-square-test . Hardy-Weinberg equilibrium (HWE) was evaluated by a goodness-of-fit Chi-square test to compare the observed genotype frequencies with the expected among the controls. Univariate and multivariate logistic regression models were used to evaluate associations between genotypes and risk of gastric cancer by odds ratios (ORs) and their 95% confidence intervals (CIs) with the adjustment for possible confounders. Risk genotypes of studied SNPs were combined to create a genetic score of the number of the observed risk genotypes, and this score was used for further analyses. All statistical analyses were performed by using Statistical Analysis Software , and all P values were two-sided with a 0.05 significance level.Differences in the distributions of demographic characteristics, selected variables, and frequencies of genotypes between cases and controls were tested by the Student's


Osteoporosis has been reported to be at least partially developed in response to functional polymorphisms of the osteoprotegerin (OPG). However, conflicting results have been found. This meta-analysis aimed to provide an assessment of the relationship between the risk for developing osteoporosis and OPG T950C polymorphism in the Chinese population. Studies to be analyzed were identified with the literature search in PubMed, Embase, Web of Science, the Cochrane Library, and the Chinese National Knowledge Infrastructure during May 2017. Seven case-control studies that included a total of 1850 osteoporosis cases and 3074 controls were assessed in this meta-analysis. Overall, no significant associations could be detected between OPG T950C polymorphism and osteoporosis when all included studies were pooled into this meta-analysis. In a subgroup analyses, OPG T950C polymorphism was significantly associated with the osteoporosis risk in South China  and for studies that included postmenopausal osteoporosis  or hospital-based controls . In conclusion, the results of this meta-analysis suggest that OPG T950C polymorphism might be associated with an increased osteoporosis risk in the Chinese population. It has previously been estimated that 30% of women and 12% of men were affected by osteoporosis . Among tWe searched for studies that examined the relationship between the OPG T950C polymorphism and osteoporosis, using the following databases: PubMed, Embase, Web of Science, the Cochrane Library, and the Chinese National Knowledge Infrastructure. The utilized keywords were (OPG or osteoprotegerin or TNFRSF11B) and osteoporosis and polymorphism and (Chinese or China) and the search was last updated in May 2017. Additional records were identified via manual search and the utilized inclusion criteria were: (1) studies that described the association between OPG T950C polymorphism and the development of osteoporosis, (2) studies that used sufficient genotypic data to calculate the odds ratio (OR), (3) all participants were of Chinese population. The following exclusion criteria were defined: (1) studies that contained incomplete data, (2) case reports, (3) editorial articles, (4) review articles, and (5) meeting abstracts.Information was carefully extracted from all eligible publications independently by two authors (Li S and Jiang H) according to the inclusion criteria. The titles and abstracts of all potentially relevant articles were screened firstly. Full articles were then scrutinized if the title and abstract were ambiguous. Disagreements were resolved through a discussion between the two authors. The following characteristics were collected for each study: name of first author, year of publication, source of controls, geographic area, sample size, and available genotypic data for OPG T950C.The relationship between OPG T950C polymorphism and the development of osteoporosis was evaluated via calculating pooled ORs, based on individual ORs. A meta-analysis examined the overall relationship of the C allele with the risk for osteoporosis development relative to the T allele; and the contrast of homozygotes CC versus TT, the contrast CC versus (TC+TT), and the contrast (CC+TC) versus TT. The heterogeneity between the utilized studies and Hardy-Weinberg equilibrium (HWE) expectations in controls were tested with a Chi-square-based Q-test . Either Eighty-eight relevant publications were identified with the outlined search strategy. In agreement with the outlined inclusion criteria, seven case-control studies –15 met oAn evaluation of the links between OPG T950C polymorphism and the risk of osteoporosis development are summarized in In order to compare the difference and evaluate the sensitivity of the meta-analyses, we used both models (the fixed-effects model and random-effects model) to evaluate the stability of the meta-analysis. The pooled results were compared between fixed-effects and random-effects models, thus evaluating the sensitivity of the meta-analysis. All significantly pooled ORs were substantially altered. Consequently, results of the sensitivity analysis suggest that the data of this meta-analysis were rather unstable .Osteoporosis is a relatively common skeletal disorder, which is characterized by low bone mass and an accompanied deterioration of the bone microarchitecture with an increased propensity to fracture . RecentlDuring the past decade, numerous published meta-analyses focused on polymorphisms of OPG T950C and a potential link to the risk for the development of osteoporosis –20. Lee Our meta-analysis has several strengths: First, we strictly followed predefined inclusion and exclusion criteria to reduce selection bias as much as possible. Second, inclusion of non-English language reports was important to minimize a major potential threat for the validity of meta-analysis-publication bias and risks related to language bias. Third, we investigated the influence of the geographic area with regard to the risk to develop osteoporosis in relation to OPG T950C polymorphism.The associated limitations should also be pointed out: First, the conducted (ethnic-specific) meta-analysis included data from a single ethnic group only; therefore, our results are only applicable to this particular ethnic group. Second, the obtained results of the sensitivity analysis were relatively unstable, which may potentially influence our obtained conclusion. Third, the etiology of osteoporosis is relatively complex and is mediated by activities of numerous genes. A single gene or a single nucleotide polymorphism might impact the osteoporosis risk less than what has been suggested so far. Finally, due to limitations of the funnel plotting (which requires a range of studies), we could not evaluate publication bias for this meta-analysis.In conclusion, this meta-analysis indicated that OPG T950C polymorphism might be linked to the risk for developing osteoporosis in the Chinese population. However, due to the limitations of this study, the presented results and conclusions should be interpreted cautiously and further large-scale studies are warranted to confirm the findings.S1 Table(DOCX)Click here for additional data file.S2 Table(DOCX)Click here for additional data file.


Punadjusted = 0.034; Dominant model: Punadjusted = 0.040); however, the result was nonsignificant when the Bonferroni correction and false discovery rate (FDR) were applied. In subgroup analyses by drinking status in the dominant model, our study revealed that the ALDH2 rs671 G>A polymorphism significantly increased the risk of GC for drinkers (dominant model: P < 0.001). No relationship between the ALDH2 rs671 G>A polymorphism and GC risk was observed in any other subgroup. Our present study indicated no association between the ALDH2 rs671 G>A polymorphism and GC risk in Eastern Asian populations. However, the ALDH2 rs671 G>A polymorphism can significantly increase GC risk for drinkers.To date, the relationship between the aldehyde dehydrogenases-2 (ALDH2) rs671 G>A (Glu504Lys) polymorphism and gastric cancer (GC) risk has not been thoroughly elucidated. To derive a more precise estimation of the effect of the ALDH2 rs671 G>A polymorphism on GC, we conducted this meta-analysis. We searched for qualified studies in the Embase, PubMed, Wang Fan and China National Knowledge Infrastructure databases. Pooled odds ratios (ORs) and 95% confidence intervals (CIs) were calculated to assess the association. A total of 6,421 GC patients and 8,832 control subjects were included in the present study. The pooled results indicated no significant relationship between the ALDH2 rs671 G>A polymorphism and GC susceptibility in all genetic models. A stratified analysis by country showed that the ALDH2 rs671 G>A polymorphism might be a risk factor for GC in Japan (Allele model:  Helicobacter pylori infection and micronutrient deficiency have been identified. Although people are exposed to the above factors, not all of them will develop GC, indicating that genetic factors are involved in the development of GC [Gastric cancer (GC) is a frequent malignant tumour and is one of the primary causes of tumour-associated deaths in the world. In 2012, 951,600 patients were diagnosed with GC, accounting for 8% of cancer cases, and 723,100 patients died of GC, accounting for 10% of cancer deaths . GastricA SNP in the aldehyde dehydrogenase-2 (ALDH2) gene, rs671, is highly prevalent among the East Asian population and causes decreased ALDH2 enzyme activity which may result in an inability to eliminate acetaldehyde. Acetaldehyde, a metabolite of alcohol, is known to increase cancer susceptibility . SeveralPrevious case-control studies have reported a correlation between the ALDH2 rs671 G>A polymorphism and the development of stomach cancer, However, the results are still discrepant , 10, 15.Figure I2 = 23.2%, Punadjusted = 0.282, PBonferroni = 1.000, PFDR = 0.461); GA VS GG ; allele model ; recessive model ; and dominant model . The heterogeneity of the included studies was low in all genetic models, and the fixed effect model was applied. The stratified analysis subgroup by country revealed an increased GC risk in Japan based on the allele model . There was no statistically significant relationship in China and Korea under any genetic models. This result suggests that the differences between countries might be a potential source of heterogeneity for this relationship. It is hypothesized that differences between countries might reflect diversity in alleles and genotypes among diverse ethnic populations. However, this result should be interpreted prudently and confirmed by more case–control studies, as the sample size in the Japan subgroup is extremely small. When the Bonferroni correction and FDR were applied to adjust for multiple comparisons, the result was unreliable and showed no association between the ALDH2 rs671 G>A polymorphism and GC risk in Japan under any models. No statistically significant relationships were found in the population-based and hospital-based subgroup according to the source of controls. When the subgroups were split up by sex, smoking status and drinking status in the dominant model, our study revealed that the ALDH2 rs671 G>A polymorphism significantly increased the risk of GC for drinkers. The impact of the ALDH2 rs671 G>A polymorphism on alcohol induced carcinogenesis has been identified in published studies of cancers of the UADT, including head and neck cancers and OSCC [When stratified by country, the subgroup study indicated that the ALDH2 rs671 G>A polymorphism significantly increased the risk of GC in Japan based on the allele model and dominant model  databases. No language limitation was used in the search and the last search was updated on July 16, 2017. The following search key terms were used:  and (‘gastric carcinoma’ or ‘gastric cancer’ or ‘stomach cancer’) and (‘polymorphisms’ or ‘genotype’ or ‘polymorphism’). We also studied the reference lists of the included studies and recent reviews.The eligible studies included in this meta-analysis must have provided the following information: (1) the relationship between the ALDH2 rs671 G>A polymorphism and GC risk; (2) a case–control study; (3) sufficient data for estimating an OR and the corresponding 95% CI; and (4) GC diagnoses and the sources of cases and controls were clearly described in the study. Studies were excluded for the following reasons: (1) duplicate data; (2) insufficient data; and (3) abstracts, reviews, comments and editorial letters. The largest published studies were selected if the same or overlapping data were used.According to the above inclusion criteria, two investigators independently extracted information from all collected studies. Discrepancies were solved by discussion among all investigators. We collected the following characteristics from each eligible study: the first author, year of publication, ethnicity, country, study design, source of cases, genotyping methods, number of cases and controls, minor allele frequency (MAF) in controls, matching variables, genotypes, source of the control group , evidence of HWE in the control group, and others.Two investigators independently assessed the quality of the included studies using the NOS. The NOS was performed to assess the study quality based on the following aspects: selection, comparability, and exposure situation in case-control studies. Rating scores ranged from 0 to 9. Studies with a score higher than seven were considered to be of good quality .P was less than 0.05. We assessed the strength of the association between the ALDH2 rs671 G>A polymorphism and GC susceptibility using ORs with 95% CIs in the co-dominant model (AA vs. GG and GA vs. GG), recessive model (AA vs. GA+GG), dominant model (GA+AA vs. GG), and allele model (A vs. G). We conducted subgroup tests according to country, source of controls, sex, smoking status and drinking status. The significance of pooled ORs was tested using Z test. Differences were considered statistically significant when P was less than 0.05. The I2 test was used to explore the heterogeneity among eligible studies [I2 value > 50% and was considered to represent significant statistical heterogeneity [I2 value < 50% and was considered to represent less heterogeneity [P < 0.05 was considered statistically significant). Funnel plots were also used to illustrate the publication bias [P-values. To adjust for multiple comparisons, the Bonferroni correction method and FDR were applied. The power of the meta-analysis for each polymorphism to detect some effect size was estimated according to the method recommended by Hedges and Pigott with P < 0.05 considered as statistically significant [HWE in the controls was assessed for each study with a goodness-of-fit test (Chi square or Fisher's exact test). HWE was considered to be in significant disequilibrium when  studies . The ranogeneity . The fixogeneity . To explion bias , 38. Allnificant .


P = 0.172). In subgroup analysis, significant decreased AD susceptibility was found among Asian population in allele model  and dominant model . Moreover, T allele of LRP1 C766T was statistically associated with late onset of AD (LOAD) . In conclusion, our meta-analysis suggested that LRP1 C766T polymorphism was associated with lower risk of AD in Asian, and could reduce LOAD risk especially. Considering some limitations of our meta-analysis, further large-scale studies should be done to reach a more comprehensive understanding.Low density lipoprotein receptor-related protein 1 (LRP1) C766T polymorphism (rs1799986) has been extensively investigated for Alzheimer’s disease (AD) susceptibility. However, results in different studies have been contradictory. Therefore, we conducted a meta-analysis containing 6455 AD cases and 6304 controls from 26 independent case–control studies to determine whether there was an association between the LRP1 C766T polymorphism and AD susceptibility. The combined analysis showed that there was no significant association between LRP1 C766T polymorphism and AD susceptibility (TT + CT versus CC: OR = 0.920, 95% CI = 0.817–1.037,  It is essentially characterised by cerebral senile plaques laden with β-amyloid peptide (Aβ), dystrophic neurites in neocortical terminal fields as well as neurofibrillary tangles of hyperphosphorylated microtubule-associated protein tau3. Besides, loss of neurons and white matter, congophilic angiopathy, inflammation, and oxidative damage are also important pathological features of AD. It is believed that genetic factors, lifestyle and environmental factors synergistically give rise to AD. Variants associated with AD have been detected in more than 20 genes, which are involved in metabolism, inflammation, synaptic activity and intracellular trafficking5.Alzheimer’s disease (AD), a progressive and lethal neurodegenerative disorder, has become a global challenge for the 21st century6. LRP1 is ubiquitously expressed in various tissues, especially high in liver, lung and brain7. In the central nervous system, LRP1 plays an important role in controlling Aβ metabolism and maintaining brain homeostasis. There are two forms of LRP1–soluble LRP1 and cell-surface LRP1. In plasma, soluble LRP1 binds to peripheral Aβ, and consequently prevents free Aβ access to the brain8. As a cell surface receptor, LRP1 can control the endocytosis of multiple ligands, mediate cell signaling transductions and regulate gene expression through its intracellular domain11. For instance, the interaction between amyloid precursor protein (APP) and cell-surface LRP1 leads to increased endosomal trafficking of APP, accelerating Aβ production. Besides that, Aβ can enter multiple cell types  through cell-surface LRP1, in which the ubiquitous apolipoprotein E (APOE) and activated alpha-2-macroglobulins (A2M) are chaperones, and subsequently degraded by endopeptidase12. Therefore, LRP1 are involved in the bulk transport, primary production, brain and systemic clearance of AD toxin Aβ, and thus plays a critical role in AD pathogenesis.Low density lipoprotein receptor-related protein 1 (LRP1) has been widely studied due to its pleiotropic roles in AD pathogenesis13. However, results in different studies have been contradictory. The inconsistency is likely to relate with insufficient statistical power, racial differences or other demographic variables. Therefore, we conducted a comprehensive meta-analysis to determine whether there was an association between the LRP1 C766T polymorphism and AD susceptibility.The silent C766T polymorphism in exon 3 of LRP1 gene (rs1799986) has attracted extensive attention since first reported as a risk factor for AD17. Finally, 26 eligible studies containing 6455 AD cases and 6304 controls were included in our meta-analysis. The characteristics of the 26 studies on LRP1 C766T polymorphism and AD susceptibility was summarized in Table A total of 167 relevant studies were identified from initial database searching, of which 35 publications were included based on titles and abstracts    .In subgroup analysis by ethnicity, T allele of LRP1 C766T was found to be associated with decreased AD susceptibility among Asian population  for heterogeneity among those studies, but gender might contributed to the heterogeneity  are co-deposited with Aβ in senile plaques in AD brains21. Subsequent studies demonstrated that LRP1 modulates the clearance of Aβ via receptor-mediated pathway in central nervous system24. Besides, soluble LRP1 provides an endogenous peripheral ‘sink’ activity for Aβ by preventing plasma free Aβ access to the brain25. It has also been reported that LRP1 is responsible for a rapid peripheral uptake of Aβ by the liver, which plays a key role in systemic clearance of Aβ26. On the other hand, endocytosis of LRP1 could modulate APP trafficking, and contribute to Aβ generation27. Interestingly, LRP1 can regulate Aβ metabolism in two contrary sides.AD, as a continuum, bring about serious threat to human health. Considering early detection and intervention at the asymptomatic stage may offer better chance of therapeutic success, it is urgent to identify early diagnostic biomarkerset al. first reported the LRP1 C766T polymorphism, and found a positive association between C allele and AD susceptibility13. This finding was replicated in some following studies30, but Kolsch et al. found the opposite result that carriers of a C allele were at lower risk of AD31, while some failed to show any association between LRP1 C766T polymorphism and AD45. Previously, three meta-analysis have tried to clarify the relationship between LRP1 C766T polymorphism and AD susceptibility, which one revealed a weak correlation of LRP1 CC genotype with AD40, but other two separately studies showed that no positive evidence was involved in the relationship between this polymorphism and AD risk among overall36 and Chinese population46. Since several factors could be responsible for these discrepancies, such as inadequate sample size, variability in phenotype definition and allele frequency polymorphisms in different ethnic backgrounds47, we conducted a comprehensive meta-analysis with different genetic models in this study, to better clarify the association between LRP1 C766T polymorphism and AD susceptibility.The association between LRP1 polymorphisms and AD susceptibility also has been described extensively, especially exon 3 C766T polymorphism. Kang 46. Compared with the results from previous studies, our data from meta-analysis was relatively reliable to illustrate the association between LRP1 C766T polymorphism and AD susceptibility, because we used different genetic models with a larger number of case-controls.New results from our research did not show any association of LRP1 C766T polymorphism with AD susceptibility from 6455 AD cases and 6304 controls in overall population. This result is consistent with two published meta-analyses13.Due to that people in different ethnic populations may have different allele frequency, and can affect the heterogeneity, we additionally conducted subgroup analysis by ethnicity, time of AD onset and APOE ε4 status. The outcomes by subgroups revealed that T allele of LRP1 C766T could reduce the risk of AD in allele model (T versus C) and dominant model (TT + CT versus CC) among Asian population, no significant role was found in Caucasian group. In terms of onset age, the results from subgroup analysis showed that T allele of LRP1 C766T could act as a protective factor for late onset of AD, but no significant association with early onset of AD. This is also consistent with previous report49. However, APOE ε4 status did not show that the influence of the association between LRP1 C766T polymorphism and AD susceptibility in our study. Moreover, our meta-regression analysis also showed that APOE ε4 status, age, and MMSE were not responsible for heterogeneity.It’s recognized that APOE ε4 is an important pathogenic factor for the development of AD. Several studies have revealed a possible protective effect of TT genotypes in carriers of APOE ε4 alleles51, MAPT52, and MAPK8IP153. In addition, some speculated that LRP1 C766T may be in linkage disequilibrium with a deleterious mutation in the LRP1 gene, or with other biologically relevant mutation on neighbouring genes, which affected LRP1 expression50. Besides, several studies have a hypothesis that the LRP1 C766T polymorphism might alter the secondary structure of the LRP mRNA to affect the translation and stability of the protein48. To date, the conclusion with LRP1 C766T polymorphism with AD susceptibility is conflicting, further genetic analyses of this locus are needed to illuminate the potential mechanism and the functional interactions with AD.LRP1 C766T polymorphism is a silent mutation, which does not change the amino acid sequence or splice site. Therefore, it is unlikely to alter the biological function by a direct causal effect with the polymorphism. Some studies consider that the LRP1 C766T polymorphism might be responsible for susceptibility to AD by interact with other genes, such as APOESome limitations of our meta-analysis should be acknowledged. The sample size in some subgroup analysis was small, which may increase the risk of false negatives or false positives. Besides, we did not perform subgroup analysis based on other factors participated in the progression of AD, such as educational background, due to a lack of sufficient information. Larger and broader independent investigations are required to better understand the role of LRP1 C766T polymorphism in AD pathogenesis.In conclusion, our meta-analysis suggested that LRP1 C766T polymorphism was associated with lower risk of AD in Asian, and could reduce LOAD risk especially. Furthermore, large-scale studies should be performed to reach more understanding of this association.We searched electronic databases PubMed, Embase and CNKI (up to August 2016) using the following keywords: (“Alzheimer’s disease” or “Alzheimer disease” or “AD”) and (“low density lipoprotein receptor-related protein 1” or “LDL receptor-related protein 1” or “LRP1”) and (“polymorphism” or “SNP” or “variant” or “genotype”) without language restriction. The bibliographies of the retrieved studies were also screened to identify relevant publications.The eligible studies had to meet all the following criteria: (1) a case–control study to evaluate the association between LRP1 C766T polymorphism and risk of AD; (2) useful data including sample size, allele or genotype distribution were given; (3) genotype distribution of controls followed the HWE. Accordingly, the exclusion criteria were as follows: (1) reviews, meta-analysis or editorial articles; (2) studies were provided with inadequate information; (3) for the studies with overlapping data, only the most relevant articles with the largest dataset were included in the final analysis.The literature retrieval and inclusion were carried out in duplication by two independent reviewers.Two reviewers independently extracted the following information: first author, year of publication, country, ethnicity, total number of cases and controls, mean age of cases and controls, proportion of female in cases and controls, AD diagnosis criteria, genotyping method, source of controls, time of AD onset, genotype or/and allele distribution in cases and controls. If conflicting results produced, two reviewers would review the publications again and reached a consensus by discussion.55), and the disagreements were judged by the third reviewer to ensure a consistent outcome. Quality scores of studies ranged from 0 (the lowest) to 15 (the highest). Studies with quality scores among 10 to 15 were grouped into high quality studies and other studies scored between 0 and 9 were categorized into low quality studies.Two reviewers independently assessed the quality of each included studies in the meta-analysis according to the criteria of quality assessment , codominant model [homozygote comparison (TT versus CC) and heterozygote comparison (CT versus CC)], dominant model (TT + CT versus CC), and recessive model (TT versus CC + CT). The heterogeneity was also quantified with I2 statistics. If no significant heterogeneity was found between the studies, the pooled OR was calculated by using the fixed effects model (the Mantel-Haenszel method)56. Otherwise, the random effects model (the DerSimonian and Laird method) was applied57. Both of univariate and multivariate meta-regression analyses were also carried out to explore potential sources of heterogeneity among studies. The log of the ORs from involved studies was using as dependent variables, and age, gender, Mini-Mental State Exam (MMSE) and/or APOE ε4 status as covariates. Publication bias was tested by Begg’s test and Egger’s test59. We also performed subgroup analysis according to ethnicity, time of AD onset and APOE ε4 status, respectively. Statistical analyses were conducted with Stata Version 11.0 , and a two-sided P < 0.05 was considered statistically significant.HWE in controls was tested by a chi-square test. Summary odds ratio (OR) with confidence interval (95% CI) for genotypes and alleles were used to evaluate the strength of association between LRP1 C766T polymorphism and AD susceptibility. The significance of the pooled OR was measured using the


Previous studies propose that hypothyroidism might play a crucial role in the pathogenesis of non-alcoholic fatty liver disease (NAFLD), but findings from published studies on the relationship between hypothyroidism and NAFLD are still controversial. Our study aimed to make a comprehensive evaluation of the relationship between hypothyroidism and NAFLD through a meta-analysis.PubMed, China Dissertation Database, and EMBASE databases were searched to find observational studies assessing the relationship between hypothyroidism and NAFLD. The pooled odds ratios (ORs) with 95% confidence intervals (95% CIs) were calculated to evaluate the strength of the relationship between hypothyroidism and NAFLD through meta-analysis.P < 0.001). Meta-analysis of 9 studies providing adjusted ORs found that hypothyroidism was independently correlated with NAFLD . Subgroup analysis found that both overt hypothyroidism and subclinical hypothyroidism were significantly correlated with NAFLD, and the pooled ORs were 1.70  and 1.40 , respectively. Besides, meta-analysis of studies providing adjusted ORs also found that both overt hypothyroidism and subclinical hypothyroidism were independently correlated with NAFLD, and the pooled ORs were 1.81  and 1.63 , respectively.Thirteen articles were ultimately included in our meta-analysis. Meta-analysis of the 13 studies found a high correlation between hypothyroidism and NAFLD (OR = 1.52, 95% CI 1.24–1.87, The meta-analysis provides strong epidemiological evidence for the relationship between hypothyroidism and NAFLD. Both individuals with subclinical and overt hypothyroidism are at higher risk for NAFLD than euthyroid subjects. The prevalence of non-alcoholic fatty liver disease (NAFLD) has increased substantially during the past decades, and it has become the leading cause of liver disease worldwide, which may be partly attributed to the rising prevalence of obesity . NAFLD iHypothyroidism is a common disease of the endocrine system that affects lifelong health. The physiological role of the thyroid gland has been taken seriously by many scholars, not just because of the critical role of thyroid hormones in cell metabolism and energy homeostasis  but alsoPrevious studies propose that hypothyroidism might play a crucial role in the pathogenesis of NAFLD. Some studies report that the prevalence of hypothyroidism is from 15.2 to 36.3% among patients with NAFLD, indicating that hypothyroidism is a common concomitant disease of NAFLD and may be related to the development of NAFLD . At presWe followed the PRISMA method for conducting a meta-analysis of observational studies . In ordeOur inclusion criteria were as follows: (1) cohort, cross-sectional, or case–control studies which investigated the association between hypothyroidism and NAFLD; (2) all studies must report odds ratios (ORs) with 95% confidence intervals (95% CIs) values or other values which could be converted into ORs; (3) included NAFLD patients must be diagnosed with an ultrasound examination or pathologic examination to make a clear definite diagnosis, and other diseases that could cause hepatic steatosis were excluded; and (4) the diagnosis of hypothyroidism must be based on biochemical tests including TSH levels and T4/FT4 levels.The final data were abstracted from each study using standardized form: the first author’s name, year of publication, study design, study location, number of participants, participant baseline characteristics (age and gender), method used to identify and verify NAFLD as well as thyroid function, definition of hypothyroidism including overt hypothyroidism or subclinical hypothyroidism. We used Newcastle Ottawa Scale to evaluate the quality of included studies . The selI2 statistic and Cochran’s Q-test, and I2 more than 50% or P of Q-test less than 0.10 showed significant heterogeneity    and NAFLD , but other covariates were not.Obvious heterogeneity was found among these 13 studies (P < 0.002)  .Meta-analysis of six studies found that overt hypothyroidism was significantly correlated with NAFLD    is the main type of thyroid hormone derivatives, and some studies has demonstrated that T2 has an lipid-lowering effect and can reduce the excess fat in cultured hepatocytes  had benefits on NAFLD in subclinical hypothyroid patients and LT4 supplementation played a beneficial role in delaying the development of NAFLD . More clIn conclusion, our meta-analysis provides strong epidemiological evidence for the significant relationship between hypothyroidism and NAFLD. Both individuals with subclinical hypothyroidism and overt hypothyroidism are at a higher risk for the development of NAFLD than those with normal thyroid function. More prospective cohort studies are needed to further strengthen the relationship between NAFLD and hypothyroidism.WH and J-aZ designed the study. WH, XA, LL, XS, QY, QL, and J-aZ contributed to the literature search, interpretation, writing, and proofreading of the manuscript. WH, XA, LL, and QY extracted data and performed data analyses.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 reviewer SP and handling Editor declared their shared affiliation.


Infant low birth weight is one of the major problems in different societies. Different reports have provided different results regarding the relationship between maternal anemia and infant low birth weight in different months of pregnancy. The aim of this study was to determine the relationship between maternal anemia during pregnancy and infant low birth weight.2 index was also used to measure heterogeneity between the studies.This systematic review was conducted using related keywords in national  and international  databases. Relative risks and confidence intervals were extracted from each study. The results were combined using random-effects model for meta-analysis. The IOverall, 17 studies with a total sample size of 245407 entered the final meta-analysis and demonstrated that the relative risk for maternal anemia in the first, second and third trimester of pregnancy were 1.26 (95% CI: 1.03-1.55), 0.97 (95% CI: 0.57-1.65), and 1.21 (95% CI: 0.84-1.76), respectively. The relationship between maternal anemia and infant low birth weight in the first trimester of pregnancy was significant. Maternal anemia, especially during the first trimester of pregnancy, can be considered as a risk factor for pregnancy outcomes. Therefore, one needs to take the necessary steps to cure this disease in order to reduce the incidence of infant low birth weight. Low birth weight (LBW) is a major problem in different societies. On average, 7.7% of infants weigh less than 2500 gr. Infants with birth weight less than 2500 gr are known as LBW infants, regardless of gestational age. According to a study, 10% of infants born in the United States are LBW infants . AccordiConsidering the high prevalence of anemia among pregnant women aged 17-45, contradictory results regarding the effect of maternal anemia on birth weight and absence of a definite conclusion in obstetrics and genealogy books, we decided to investigate this topic using meta-analysis methods , 11. TheData sourceseither/or” structure was applied for further searching; references were also examined to find further necessary sources on the subject. This study was a systematic review of the literature and methods based on PRISMA guideline from the beginning of 1990 until April 2017 . Valid kStudy selection (Inclusion and exclusion criteria)STROBE, which is a standard and famous international checklist, was used to determine the quality of the study . This chStudy selectionFirst, researchers prepared a list of topics and abstracts of each study at the beginning of study and studies that covered the relationship between maternal anemia and LBW based on pregnancy month entered the main study. Cohort, case-control and cross-sectional that evaluated the relationship between maternal anemia and LBW were analyzed in the present study. Relative risk with 95% confidence interval was used to assess the significance of the study.Data extractionData extraction was performed by two researchers independently and data extraction form  was used to minimize bias and error in collecting data. In cases of necessity, ambiguity of article information and some certain questions, we contacted the authors by E-mail. The data extracted by the two researchers was compared and discussed in case of possible discrepancies; the dispute was shared with a third person and eventual consensus was reached through re-examination of the issue. Risk of bias assessmentOne of the risks of bias assessment was the fact that data extraction was conducted by two researchers. There was also the risk of dispute.OutcomesofinterestLBW defined as an infant born at less than 2,500 gr. And hemoglobin level in the first and the second trimester of pregnancy is less than 11 g/dL and is less than 10.5 g/dL in the third trimester, this condition is considered anemia .Statisticalanalysis2 index and Q test were used to assess the heterogeneity. Considering the heterogeneity of the studies, random effects model was used to combine different results. STATA Ver 3.2 was used for data analysis and p<0.05 was considered significant for all analyses. In present study, the extracted data included study design, type of the study, name of the author, year of publication, continent, number of participants, trimester and the outcome of pregnancy (LBW). The level of hemoglobin was classified in three groups of <14, 11-13, 11> g/dL, while <11 was considered maternal anemia. LBW, as an outcome of pregnancy, was analyzed when the infant weighed less than 2500 grams. The effect of hemoglobin was evaluated using relative risk index (RRI) in three different phases  of pregnancy based on the weight of the infant. IThe minimum and maximum relative risks (RR) or odds ratio (OR) at upper and lower limits of 95% confidence intervals were extracted for each study. In studies, for which this index was not extracted (articles without the OR to use the formula), the index was calculated using OR=ad/bc formula and OR or RR logarithm were used to symmetrize the effect size. Error standard was calculated using the following formula: SE= 1n (upper 95% CI /lower 95% CI) /(2×1.96). If only upper and lower limit had been reported in studies, OR= ad/bc was used. Meta-regression was used to investigate the relationship between the year of study, effect size and causes of heterogeneity. Beggs Funnel Plot was used to examine the publication bias. Sensitivity analysis was used to assess the effect of each study on the overall result.To analyze sensitivity, the effect of each article was excluded and the result was combined with other studies. This shows whether exclusion of this study has significant effect on the overall result or not. In the case of our study, the effect was not statistically significant.Overall, 2454 relevant articles were identified by searching databases. Studies that did not yield any helpful and relevant information were excluded; 76% of them were cohort studies with an age range of less than 30 years old. 70% of articles were conducted in Asia. , Table IThe relationship between maternal anemia in the first trimester of pregnancy and LBWBy type of studyAs illustrated in It seems that Hamalainen’s study has estimated this relationship with relatively high risk . And The2 index showed that the results of 9 cohort studies were heterogeneous and this heterogeneity was significant . The heterogeneity of 2 case-control studies was 0.0% and this heterogeneity was not significant (p=0.481). Ov each article was measured individually and it was found that their weight is calculated when the data is combined. Among cohort studies, Levy’s study had the highest weight because of large sample size  (According =0.0061) , 29, 30.Relationship between maternal anemia and LBW in the third trimester2 index between this studies was 72.8%, which shows a significant heterogeneous relationship between them (p=0.001)  .The present study is the first systematic review and meta-analysis research about the relationship between maternal anemia and LBW. Different studies with different values of relative risk and confidence interval were previously conducted , 28-33. et al  Inability of databases to search using combined keywords Some studies were excluded due to low quality and insufficient proofsExclusion of some studies due to ambiguity in the month of pregnancyDifferent definitions for anemia in different studiesExclusion of some studies and articles due to improper direction of publications and lack of causative relationship. Considering the relationship between maternal anemia and LBW in the first trimester of pregnancy, anemia can be used as a preventive factor for infant LBW. Therefore, we need to cure maternal anemia to prevent or cure LBW.