Background: Arsenic is a toxic metalloid element widely distributed throughout the environment. Arsenic contaminated water has become an ongoing public health issue affecting hundred million people worldwide. The aim of this paper was to summarize the evidence in the association between arsenic metabolites and urinary tract cancer risk. Methods: A systematic review was conducted searching for observational studies that evaluated the association of arsenic metabolites and urinary tract cancer. Risk estimates from individual studies were pooled by using random effects models. Results: All the metabolites considered in this study resulted to be significantly associated to urothelial cancer, respectively: IA% 3.51 (1.21–5.82) (p = 0.003), MMA with WMD = 2.77 (1.67–3.87) (p < 0.001) and DMA with WMD = −4.56 (−7.91–1.22) (p = 0.008). Conclusions: Arsenic metabolites are significantly associated to urothelial cancer. Future studies will help to verify the independent association(s) between arsenic metabolites and urothelial cancer. Arsenic (As) is a toxic metalloid element widely distributed throughout the environment. It can be found in rocks, soil, water, air, or in plants and animals in organic or inorganic forms. Naturally high levels of arsenic are present in the groundwater of a number of countries  ,2. ArsenReview QuestionThe aim of this paper was to summarize the evidence in the association between As metabolites and urinary tract cancer risk provided by up-to-date research outcomes.Mean Difference (MD) (cases vs. controls) of IA (%);Mean Difference (MD) (cases vs. controls) of DMA (%);Mean Difference (MD) (cases vs. controls) of MMA (%).The primary outcomes evaluated were: A sensitivity analysis was conducted on odds ratios reported in the included papers about the risk of urothelial cancer of cases vs. controls. In this case, the exposure of IA was classified as high or low (not continuous). The cutoffs to define as high/low IA% changed between the studies. The systematic review was conducted according to the “Conducting Systematic Reviews and MetaAnalyses of Observational Studies of Etiology” (COSMOS-E) guidelines . Only obTwo authors, P. Scampoli and F. Meo, independently screened titles and abstracts of retrieved citation in order to identify potentially eligible studies. Each discrepancy was discussed with a third investigator, T. Staniscia. All potentially relevant citations were then retrieved in full-text and independently reviewed by two investigators, F. Cedrone and G. Di Martino Data meeting the selection criteria were independently extracted from articles by two investigators, G. Di Martino and P. Scampoli, and reported in a specific digital form. The variables considered for this systematic review were: study name, year, country, study design, number of participants, mean of IA%, MMA% and DMA% among cases and controls, risk estimates, factors considered in the adjusted analysis and conclusions. All the extracted data were checked by two investigators, P. Di Giovanni and F. Romano. Two investigators (F. Cedrone and F. Meo) assessed the study quality according to the Newcastle-Ottawa Quality Control Scale .p-values was < 0.10, we pooled risk estimates using random effects models. The heterogeneity was also represented by I2 index. The potential for publication bias was assessed using a funnel plot in conjunction with Kendall’s Tau, unweighted and weighted Weighted Egger Intercept, Macskill’s inverse variance weighted and Macskill’s inverse pooled variance weighting. All analyses were carried out using R version 3.2.0.3.The characteristics of included studies were showed in Electronic searching through 14 March 2020 retrieved 9659 articles, as shown in Characteristics of included studies are shown in p = 0.003), MMA with WMD = 2.77 (1.67–3.87) (p < 0.001) and DMA with WMD = −4.56 (−7.91–1.22) (p = 0.008) as shown in p < 0.001), a higher concentration of MMA had overall OR of 1.44 , and a higher concentration of DMA had overall OR of 0.43 . The study by Chung et al. [p-values was <0.10 for all outcomes and for this reasons we pooled risk estimates using random effects models. Every publication bias method suggests the absence of publication bias  was performed in the same country (Taiwan), and it can make these results less generalizable. From included studies is not possible to find a threshold for the carcinogenic risk of As metabolites. So public health decision-makers need focused prospective studies both to evaluate the minimum level of risky As metabolism concentrations, both to evaluate the risk of long time exposure to As. However, the consistency of these findings sets the stage for additional research to determine the causality of these associations. Arsenic metabolites levels in urine resulted significantly associated with urothelial cancer. The presence of As in soil and drinking water is a public health concern due to its possible role in urinary tract carcinogenesis. Public health policies and intervention should be aimed to improve surveillance and preventive strategies and to protect population exposed to arsenic. Future studies using large sample size, appropriate baseline, and prospective arsenic metabolism estimation will help to verify the independent association between arsenic metabolites and UC. In addition, a prospective evaluation of chronic exposure and As metabolism over time, could help in assessing the dose-response effect on urinary tract and its impact on UC development.


According to several studies, the autoimmune response may lead to osteoarthritis and dyslipidemia and may affect the homeostasis of the human body's internal environment and then cause its own immune regulation. Consequently, the risk of osteoarthritis might be increased by dyslipidemia, but this association is not universally acknowledged. Therefore, a systematic review and meta-analysis was conducted to study the relationship between dyslipidemia and the risk of osteoarthritis.  In this study, PubMed, EMBASE, and the ISI Web of Science were used to identify related studies published before July 2018. The relationship between dyslipidemia and the risk of osteoarthritis was evaluated on the basis of relative risk (RR) values and the corresponding 95% confidence intervals (CIs). To further investigate this relationship, we also employed the random effects model proposed by DerSimonian and Laird.  A total of nine studies were included to study the effect of dyslipidemia on the risk of osteoarthritis, including four cohort, three case-control, and two cross-sectional studies. Among these studies, six stated data for knee osteoarthritis, two reported on hand osteoarthritis, and one reported on hip osteoarthritis. A total of 53,955 participants were included in the meta-analysis, comprising 22,501 patients with OA . Based on the meta-analysis of case-control and cross-sectional studies, osteoarthritis was clearly higher in those with dyslipidemia compared to those who did not suffer from dyslipidemia . In addition, the meta-analysis of cohort studies did not present any relationship between dyslipidemia and OA .  Even though our meta-analysis of case-control and cross-sectional studies suggested a strong relationship between dyslipidemia and osteoarthritis; this relationship was not validated by our meta-analysis of only cohort studies. As a result, further investigation needs to be conducted on the relationship between dyslipidemia and osteoarthritis, considering the significant public health relevance of the topic. Osteoarthritis (OA) refers to a chronic degenerative disease that involves the cartilage, as well as its surrounding tissues . OA is cThis research was performed according to the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) Statement  and the Published studies in PubMed, EMBASE, and the Web of Science were searched based on the following keywords: (“hyperlipidaemia” OR “dyslipidemia” OR “triglyceride” OR “cholesterol” OR “lipoprotein” OR “lipid” OR “metabolic syndrome”) and (“OA”). No restrictions on language or the date of publication were placed. Additionally, this study also searched the reference lists. Unpublished studies and original data were not included.The eligibility criteria were as follows: study design ; an exposure factor of blood lipid levels and an outcome of OA; availability of the odds ratio (OR)/risk ratio (RR) values and corresponding 95% confidence intervals (CIs) for dyslipidemia patients and the general population; or the availability of sufficient information to measure these variables. The most recent all-inclusive study was searched under the condition that two studies used the same population. The definition of dyslipidemia was in line with the US National Cholesterol Education Program Adult Treatment Panel III guidelines. In accordance with the National Cholesterol Education Program, the definition of dyslipidemia was high-density lipoprotein cholesterol (HDL‐C) < 40 mg/dL, as well as total cholesterol, low-density lipoprotein cholesterol (LDL-C), and TG levels of ≥200, ≥130, and ≥130 mg/dL, respectively . The ACRTwo scholars (J.X. and J.L.) obtained the essential information from the chosen studies according to the standard. The following information was gathered: name of the first author, publication year, country in which the research was carried out, study design, number of participants, period of follow-up, sources of controls, potential adjusted confounding variables, OR/RR values, and 95% CIs.To date, no available common scale has been proposed to evaluate the quality of all kinds of observational studies. As a result, two authors individually employ the modified Newcastle-Ottawa Scale (NOS)  as reporI2 statistic was employed to evaluate heterogeneity between the studies. Low, medium, and high heterogeneities were categorized as 25%, 50%, and 75%, respectively [p value was less than 0.1. Sensitivity analyses were conducted by altering the pooling model [n < 10) [The random effects model put forward by DerSimonian and Laird was applied to investigate the relationship between dyslipidemia and the risk of OA among the cohort studies . The I2 ectively . Definitng model . In adding model  and Eggeng model  tests. N[n < 10) .Furthermore, we conducted a meta-analysis of the case-control and cross-sectional studies in regard to the influence of dyslipidemia on the risk of OA and expressed the results as pooled risk ratios with 95% CIs with the application of a random effects model.STATA version 12.0 (Stata) was carried out to perform all statistical analyses.The process of study selection for the meta-analysis can be found in I2 = 0%) (n < 10). Nevertheless, Begg's (p = 0.697) and Egger's (p = 0.465) tests failed to identify substantial publication bias.Reports from four studies permitted the calculation of effect estimates for OA –36, 38. I2 = 0%) . In the I2 = 0%) . No testI2 = 0%)  . Based oI2 = 0%)  .Admittedly, this is the first meta-analysis to investigate the relationship between dyslipidemia and the risk of OA. Even though a meta-analysis of cohort studies did not present any connection between dyslipidemia and OA, a meta-analysis of case-control and cross-sectional studies demonstrated that the risk of OA was obviously higher among those suffering from dyslipidemia compared to those without dyslipidemia.The potential pathophysiological mechanisms accounting for these results remain unknown. Changed expression of cholesterol influx genes in human osteoarthritic chondrocytes and in the cartilage of patients with OA has been depicted . MoreoveThere are several strengths of our study. First, this is the first meta-analysis to investigate the relationship between dyslipidemia and the risk of OA with a large sample size , probably providing a reference in regard to this relationship. Clinically, the results indicate that patients with dyslipidemia are required to focus more on their risk of OA in comparison with the general population. Second, a comprehensive literature search was conducted using PubMed, EMBASE, and the Web of Science to recognize potential studies from which to examine the relationship between dyslipidemia and the risk of OA. Additionally, most of the studies included were high-quality studies. All of these features enhance the reliability of our study.However, this study still has some limitations. First, a meta-analysis of the different locations of OA, such as the knee, hip, and hand, was not conducted due to the limited number of included studies. The effect of dyslipidemia on the risk of knee vs. hip OA may have differences because of the diverse susceptibility of these two joints to metabolic factors. The knee may have more dependence on soft tissue and neuromuscular control for its stability. Comparatively, in the hip, the bony shape and joint congruence seem to have greater significance on the development of hip OA, which makes the hip less vulnerable to the effects of inflammation. Second, the observed outcome was an association, which is considered a confounding bias. Seven studies reported data on knee or hip osteoarthritis, but not all of the studies were adjusted for BMI or physical activity. Zhou et al. , Xie et In conclusion, considering there was a lack of evidence from the cohort studies but that strong connections were noted in the case-control and cross-sectional studies, the current systematic review and meta-analysis offers inadequate but fascinating evidence of the effects of dyslipidemia on the risk of OA. In general, the findings indicate that dyslipidemia might exert a significant pathogenic role in the development of OA, as well as offers a rationale for the shared care of patients by metabolic physicians. To be specific, structured clinical trials with predefined criteria for patient selection are still required to investigate the role of lipid-lowering therapies in the management of OA.


MTHFR gene C677T (rs1801133) polymorphism and alcohol dependence (AD), but the results of this association remain conflicting. Accordingly, authors conducted a meta-analysis to further investigate such an association. PubMed, Elsevier Science Direct and Springer Link databases were searched for studies on the association between the MTHFRC677T polymorphism and AD. Pooled odds ratio (OR) with 95% confidence interval (CI) was calculated using the fixed- or random-effects model. Statistical analysis was performed with the software program MetaAnayst and MIX.A total of 11 articles were identified through a search of electronic databases, up to February 28, 2020. The results of the present meta-analysis did not show any association between MTHFR C677T polymorphisms and AD risk . Results of subgroup analysis showed no significant association between MTHFR C677T polymorphism with AD in Asian as well as in Caucasian population. In conclusion, C677T polymorphism is not a risk factor for alcohol dependence.Alcohol dependence is a complex neuropsychiatric disorder. Numerous studies investigated association between  MTHFR gene is present on chromosome 1p36.3. Numerous single nucleotide polymorphisms (SNP) are known in MTHFR gene like C677T and A1298C etc MTHFR gene C677T polymorphism with AD. However, findings showed inconsistent results Homocysteine is a sulfur containing amino acid, several genetic and environmental risk factor are reported for higher plasma concentration of homocysteine 2.Present meta-analysis is carried out according to MOOSE  guidelines.2.1.MTHFR” or “C677T” or “rs1801133” or “polymorphism” and “Alcohol dependence” or “Alcoholism” or “AD” or “Addiction”.Articles were retrieved through Pubmed, Google scholar, Springer Link, and Science Direct databases up to February 28, 2020, using following key words: “Methylenetetrahydrofolate reductase” or “2.2.MTHFR C677T polymorphism and alcohol dependence association was investigated in the study, (2) MTHFR C677T genotype/allele numbers in alcohol dependence cases and controls were given in the study, (3) sufficient information for calculating the odds ratio (OR) with 95% confidence interval (CI) and (4) Articles published in English language were only considered. Major reasons for studies exclusion were as follows: (1) no alcohol dependence cases analyzed, (2) the C677T polymorphism details information missing, and (3) duplicate article.Inclusion criteria were following: (1) 2.3.Name of first author, country name, number of cases and controls, number of genotypes in cases and controls and journal name with full reference from each article were extracted.2.4.2 statistic All analysis were done according to the method of Rai et al. 3.3.1.Selection of studies is given in fow diagram . Followi3.2.MTHFR C677T polymorphism. The prevalence of C and T alleles in AD cases was 71.22% and 28.79% respectively. The percentage frequency of TT genotype among cases and controls was 9.43% and 12.98%, respectively whereas prevalence of CT heterozygote among AD cases was 38.72% and 39.21% in controls. The prevalence of CC homozygote among AD cases and controls was 51.85% and 47.80%, respectively. Genotypes were in Hardy-Weinberg equilibrium in all controls. In control group the percentage of C and T allele frequencies was 67.41% and 32.59% respectively : OR = 1.04, 95% CI = 0.88–1.24; CT vs. CC (co-dominant): OR = 1.02, 95% CI = 0.62–1.68; for TT+CT vs. CC (dominant): OR = 1.10, 95% CI = 0.94–1.29; for TT vs. CC (homozygote): OR = 1.01, 95% CI = 0.66–1.51; for TT vs. CT + CC (recessive): OR = 0.97, 95% CI = 0.66–1.40) (No significant association was observed between the 66–1.40) .heterogeneity = 0.02, Q = 20.64, I2 = 51.56%, t2 = 0.04, z = 0.69), co-dominant genotype , homozygote genotype , and recessive genotype  comparisons. The “I2” value of more than 50% shows high level of true heterogeneity.A true heterogeneity existed between studies for allele contrast ; and Caucasian population    and 5.3.5.Symmetrical shape of Funnel plots' revealed absence of publication bias. P values of Egger's test were more than 0.05, also provided statistical evidence for the funnel plots' symmetry  .4.in vitro studies has demonstrated that homocysteine has neurotoxic effects especially on dopamine neurons of reward pathway MTHFR C677T genotype and AD risk in two ethnic populations, including over 3,205 subjects, our meta-analysis provides an evidence that TT and CT genotypes or T allele are not associated with AD risk. Hence the present meta-analysis indicated that C677T is not a risk factor of AD.In vivo and Meta-analysis is a statistical tool to combine the information of independent case-control studies with similar target Several limitations that should be acknowledged like (i) calculated crude Odds ratio, (ii) included the less number of available studies (10 studies) and the limited sample size of each included study, (iii) observed higher between study heterogeneity, (iv) considered only one gene polymorphism and (v) not considered other confounding factors like diet, gender etc. In addition to limitations, current meta-analysis has several strength also such as—higher study power and larger sample size in comparison to individual case control studies, and absence of publication bias etc.MTHFR 677TT genotype is not a risk factor for AD. The results of present meta-analysis should be interpreted with certain cautions due to presence of higher heterogeneity and small number of included studies. Future large-scale, population-based association studies from different regions of the world are required to investigate potential gene-gene and gene-environment interactions involving the MTHFR C677T polymorphism in determining AD risk.In conclusion, pooled analysis of data from 11 separate articles indicates that the


Compared with non-MetS, MetS was significantly associated with high CAD risk . The MetS components were also significantly correlated with high CAD risk . The presence of two , three , four , or all five MetS components , were significantly associated with a high risk of CAD. MetS and its individual or combined elements were linked with high CAD risk based on contemporary evidence. Thus, the assessment of MetS and its components might help identify people at a higher risk of advancing CAD in the future.Although numerous studies have described the link between metabolic syndrome (MetS) and Coronary Artery Disease (CAD), no meta-analysis has been carried out on this relationship. Thus, the present study intended to address this limitation. A systematic search was carried out using electronic databases, such as PubMed, CINAHL Plus, Medline, and Web of Science. A sum of 10 studies ( The World Health Organization (WHO), the European Group for the study of insulin resistance, the 2005 National Cholesterol Education Program (NCEP)-Adult Treatment Panel III (ATP-III), and the 2006 International Diabetes Federation formulated the definition of Metabolic Syndrome (MetS) after great debate ,2. MetS The enhanced pervasiveness and degree of MetS and its elements are causing harmful effects on coronary circulation . HoweverWhile differently accepted definitions have been used in the published observational studies, the current meta-analysis of these studies tested the association of MetS and its segments with CAD risk. The primary hypothesis is that MetS would associate significantly with CAD risk. Furthermore, MetS components both individually and in combination would significantly associate with CAD risk.The current meta-analysis was conducted based on a recommendation by the Meta-analysis of Observational Studies in Epidemiology (MOOSE) group . LiteratAs advised by MOOSE , the seaThe studies involved met the following inclusion criteria. The observational studies used for this work included MetS and its components as exposure and CAD as an outcome in adults aged 19 years and more. MetS components were as follows: (1) obesity or high WC, (2) advanced BP, (3) low HDL-c, (4) raised triglycerides (TG), and (5) raised hyperglycemia. MetS and its composition has been defined elsewhere . Patients who have not experienced a recent acute incident of a known cardiovascular disease are defined as CAD . CAD is Data were individually filtered by extracting potentially relevant studies based on the title and abstract. According to the eligibility criteria, the relevant studies were selected separately by reading the full-text. The final eligible studies were then chosen according to the inclusion and exclusion standards.In this meta-analysis, we included prior studies that reported odds ratios (ORs) and 95% confidence interval (CI) for the association of MetS and its segments to CAD risk compared to non-MetS. The exclusion criteria, OR and CI participants with a diagnosis other than MetS and its components were not reported. Studies that did not explore the link between MetS, its components, and CAD risk were eliminated.The primary outcome was the OR of CAD in people with MetS compared to non-MetS subjects. The secondary outcome was the OR of CAD in individuals with any element of MetS. The territory outcome was the OR of CAD in individuals with two or more MetS segments. A satisfactory agreement was defined as adjusting at least two of six covariates, including age, sex, family records of CAD, body mass index (BMI), low-density cholesterol, and smoking status. 2 statistic was applied to test heterogeneity. The heterogeneity between studies was interpreted as moderate (30%–59%), substantial (60%–89%), or considerable (90%–100%) [p < 0.05 was described as of statistical importance. We scrutinized multivariate-adjusted outcome data, displayed as OR and 95% CI. The reversed variation method was applied for investigation after log-transformation of study-specific assessments. For example, 95% CI was converted into standard error (SE) employing the following equation: SE = upper CI - lower CI/3.92. The I0%–100%) . Studies0%–100%) . All ana0%–100%) . The p <n = 3048 and women, n = 2310 were controls. The average age of all participants was above 45 years.Out of 1268 studies identified in a database search, a total of 10 studies ,29,30,312 = 13%, p = 0.33). MetS was significantly associated with high CAD risk .2 = 67%, p = 0.02). 2 = 79%, p = 0.003), high TG , and high BMI  groups, within the six studies. Six studies ,26,28,302 = 83%, p = 0.01). Significant heterogeneity was found among all studies that reported ORs of CAD with any component of MetS . Thus, any MetS component was correlated significantly with high CAD risk .Five studies ,26,28,30p = 0.004). Thus, two MetS components’ presence was significantly associated with high CAD risk .p = 0.001). The presence of three MetS components was significantly associated with high CAD risk . Six studies [p = 0.04). The presence of four MetS components was significantly associated with high CAD risk .Six studies ,26,28,30 studies ,26,28,30p = 0.25). The presence of all five MetS components was significantly associated with high CAD risk .Four studies ,26,28,30There is no bias between the studies in the funnel plot associated with MetS, any of the five MetS, and CAD . The assThe current meta-analysis examined the association of MetS and its five elements with CAD risk among adult populations. The findings revealed that MetS correlated significantly with high CAD risk compared to non-MetS. Furthermore, any individual MetS element was significantly linked with high CAD risk. The attendance of two or more MetS components was also connected with high CAD risk compared to the absence of any component. The presence of from two to four MetS components significantly increased the correlation in the severity of CAD. To our understanding, this is the first meta-analysis that presented evidence on the association of MetS and its elements with CAD risk. A longitudinal study concentrated on the consequences of MetS on BP control and its function in stimulation of arterial aging in hypertensive treated people . The resMetS is likely to result in the non-alcoholic fatty liver condition, T2DM, and vascular conditions such as stroke, peripheral vascular disease, and CVD . InitialThere were no randomized control trials (RCTs) observed during the literature search on this subject. The majority of the published retrospective studies fitted the inclusion standards added to this meta-analysis ,25,28,31The relation between MetS and CAD risk has been noted in individual studies among various populations aged 50 years or above. These studies vary considerably in their methodology and MetS definition. Various stratified analyses were conducted to examine possible heterogeneity causes, and the outcomes helped our overall conclusions. The sign of a relationship between MetS and CAD risk appeared with no significant heterogeneity between the studies.Our findings are comparable with preliminary proof regarding MetS elements. Waist circumference , elevateIn the meta-analysis, as mentioned earlier, the association of two or more MetS components with CAD risk was similar to an earlier study . That coThis meta-analysis’ major strength included studies that defined MetS well, along with its elements , and CAD. Another strength of this investigation is the large sample size of 9327 participants from 10 incorporated studies. However, some limitations have been acknowledged when understanding the results of this meta-analysis. First, there were no prospective studies on this topic. This kind of evidence may give more precise evaluations than retrospective studies, because they usually have more reliable controls for possible confounders, including socioeconomic, lifestyle, and health factors related to MetS progress. They also have adequate continuance of follow-up to identify MetS cases.Moreover, these studies collect susceptibility information before the appearance of diseases. Therefore, the temporal causal relationship can be examined. Secondly, subgroup analyses, such as sex (men versus women) and follow-up duration, were not performed due to the lack of data concerning sex and variation in the study design. Thirdly, inter-study variations in MetS definition and patients’ demographics might also influence the results. Fourthly, the studies that did not show output in OR and 95% CI were excluded. This might decrease the validity of this study. Overall, this is an observational meta-analysis study showing the association, but not the interconnection significance.The current findings are significant for public health since the ubiquity and frequency of MetS have rapidly increased globally . A possiThe relationship between MetS, all five components of MetS, and CAD indicates a symmetrical scatter pattern with no publication bias. A potential reaction could be a fixed-effect analysis that assumes little or low heterogeneity and a single underlying effect across studies . AsymmetIn conclusion, the present meta-analysis intended to examine the association of MetS and its five elements with CAD risk. This meta-analysis shows that MetS and its elements are significantly associated with high CAD risk. The initial evaluation of MetS and its five components might help identify individuals at higher risk of CAD. The current findings suggest that MetS and its components can be tested to validate CAD risk. Future study is required for establishing high-quality evidence by including RCTs or prospective cohort studies in the identification of the link between MetS and CAD risk among the global population.


Q statistic and I2 statistic), a fixed effect model  or a random effects model (for heterogeneity) was used to calculate the pooled odds ratios (ORs). We explored the potential dose-response relationship between smoking and NIHL as well. In total, 27 studies involving 30,465 participants were included. Compared with non-smokers, the pooled OR of current smokers was 2.05 : 1.71–2.46), and of former smokers was 1.11 (95% CI: 1.05–1.18). We found a curve linear association between an increasing number of pack-years (packages/day × smoking years) and risk of NIHL. The dose-response meta-analysis suggested that when the number of pack-years was less than fifteen, the risk of NIHL was increasing, and the highest combined OR was 5.25 (95% CI: 2.30–11.96) for pack-years of fifteen. After fifteen pack-years, the pooled OR had a slow decline. Our study indicated that smoking is a risk factor for NIHL. Current smokers have a higher risk than former smokers, and there is a positive dose-response relationship between smoking and NIHL.The purpose of this study was to synthesize the results of previously published observational studies through meta-analysis to clarify the association between smoking and noise-induced hearing loss (NIHL). We searched several databases as of October 2019. Based on the results of heterogeneity analysis ( Noise-induced hearing loss (NIHL) is chronic and irreversible sensorineural hearing loss resulting from long-term exposure to noise. It affects the daily social life of patients and brings a huge burden on society and economy. According to the WHO, around 466 million people worldwide suffer from disabling hearing loss, and it is estimated that unaddressed hearing loss poses an annual global cost of US$750 billion . About 1NIHL is related to multiple factors. In addition to occupational noise, other factors  the study design was a cohort, case-control or cross-sectional; (2) study population had a history of occupational noise exposure; (3) NIHL was clearly defined as the outcome; (4) study provided odds ratio (OR) or relative risk (RR) with the corresponding 95% confidence interval (CIs). If a study was published in multiple papers, we included only one with sufficient information. Review, conference, or experimental articles were excluded. Two investigators (X.L. and X.R.) independently extracted the following information from eligible studies: first author, year of publication, country, source of participants, study design, sample size, age, gender, diagnostic of NIHL, smoking information, adjusted OR/RR with 95% CIs, and adjusted or matched variables. Disagreements were solved through discussion.We used the Newcastle-Ottawa Scale  to assesOR was used as a measure of the association between smoking and NIHL. Due to the low incidence of NIHL, the reported RR was approximately considered as OR. When smoking status was just divided into smokers and non-smokers, smokers were defined as current smokers. In addition, when OR was reported separately at different smoking levels, we extracted the highest level of results. Two articles ,20 separQ test and I2 statistic to quantify the heterogeneity of studies. If P value for Q test was more than 0.10 and I2 value was less than 50%, we used a fixed effect model. Otherwise, a random effects model was applied [Before calculating the overall pooled OR, we used  applied . And we In addition, a dose-response analysis of pack-years and NIHL was estimated. Pack-years is a measure of the amount of cigarettes a person has smoked over an extended period, which is equal to the number of packages (/20 cigarettes) per day multiplied by the smoking time (/years). For example, smoking one package every day for two years equals to two pack-years. Articles that provided at least three quantitative categories were included in this calculation. Since all studies reported dose in groups, we assigned the midpoint of the group range as the dose value, and for the highest open-ended group, multiplied the lower limit by 1.5 times. We evaluated the potential curve liner relationship between the number of pack-years and NIHL by using restricted cubic splines with three knots  of the distribution ,23. We tP-value < 0.05 was considered statistically significant. A sensitivity analysis was also conducted to see the influence on the overall result by omitting each study. In addition, we recalculated the pooled OR after omitting those articles with extremely high ORs (>10). We used Begg’s rank correlation test and Egger’s linear regression test to assess potential publication bias ,26. If pBased on the database and search terms, we obtained 1541 articles to screen. Through browsing titles and abstracts of papers, we excluded 1467 records. After retrieving and reviewing 74 full articles, we excluded 47 records. Seven studies were excluded because participants in the research didn’t have a history of occupational noise exposure. A further 30 studies were excluded because they didn’t provide effect size estimate and 95% CIs to calculate the pooled OR. 4 studies were excluded because they used auditory threshold as outcome. Two studies were excluded because results were repeatedly reported in other articles. Four studies not published in Chinese or English were also excluded. Our meta-analysis includes 27 studies ,45,46,47Q test P < 0.001, I2 = 87%).Of the 27 included articles, 29 studies (2 articles reported OR separately for different noise exposure history) assessed the association between current smokers and risk of NIHL. Among the 29 studies, 20 reported a positive relationship between current and risk of NIHL, while nine found no association. Q test P = 0.394, I2 = 4%).Six studies (1 articles reported OR separately for different noise exposure history) provided information on the association between former smokers and risk of NIHL. Q test P < 0.001, I2 = 89%), and for cohort studies and case-control studies, there was no evidence of heterogeneity . According to gender, the pooled OR was 3.05 (95% CI: 1.90–4.89) for male, 1.50 (95% CI: 1.28–1.76) for both, and 1.52 (95% CI: 1.03–2.27) for female. According to race, the pooled OR was 1.88 (95% CI: 1.50–2.36) for Mongoloid, 2.41 (95% CI: 1.70–3.42) for Caucasian, and 1.52 (95% CI: 1.03–2.27) for others. According to the number of adjusting variables, the pooled OR was 1.58 (95% CI: 0.86–2.90) for 0, and 2.18 (95% CI: 1.77–2.69) for ≥1.P < 0.001). There was a non-linear increase in risk of NIHL with increasing of pack-years (about or less than 15). Compared with non-smokers, the estimated ORs of NIHL were 2.53 (95% CI: 1.58–4.05) for five pack-years of cigarette intake, 4.34 (95% CI: 2.07–9.11) for 10 pack-years, 5.25 (95% CI: 2.30–11.96) for 15 pack-years, 5.10 (95% CI: 2.31–11.27) for 20 pack-years, 4.56 (95%CI: 2.22–9.38) for 25 pack-years, 4.06 (95% CI: 2.13–7.77) for 30 pack-years, and 3.62 (95% CI: 2.03–6.46) for 35 pack-years. We found a substantial heterogeneity across studies .For the dose-response analysis, a total of 8 studies were included based on our inclusion criteria. The 8 included studies reported data for pack-years of cigarette intake and risk of NIHL. As The results were not significantly different after omitting two studies ,42 with P = 0.007; P < 0.001). In view of this, we used trim-and-fill method to recalculate the pooled OR. The result was 1.34 (95% CI: 1.10–1.64), which still indicated the same positive association. Publication bias about former smokers and NIHL was not found by either the Begg’s test or the Egger’s test .In this meta-analysis, we confirmed the hypothesis that smoking is associated with increased risk of NIHL. Both current smokers and former smokers had a higher risk of NIHL than non-smokers. And we found a dose-response relationship between smoking and NIHL.However, the specific mechanism of smoking and NIHL is unclear. Nicotine and other substances in tobacco may have ototoxicity, damaging cochlear hair cells by increasing carbon monoxide hemoglobin or reducing the volume of cochlear blood flow ,49. In aOur study showed that the pooled OR for current smokers was higher than OR for former smokers, indicating that quitting smoking could reduce the risk of NIHL. It may be associated with the dose-response relationship between smoking and the risk of NIHL risk; former smokers have lower exposure to smoke. We performed a dose-response analysis based on eight eligible studies. It showed a non-linear relationship between cigarette intake pack-years and NIHL. The OR and its 95%CI of each dose are always greater than one. The general trend also showed that the risk of NIHL would increase with the smoking dose increasing. The slow decline in the fitting model may be due to the fact that the dose concentration of the included studies was mainly between 0 and 15 pack-years. In the total of 19 dose points, only three points were larger than 15, which made the latter trend less accurate. In addition, the studies included in the dose-response analysis were mainly cross-sectional studies. There might be a healthy worker effect in the population. Therefore, a smoking cessation program is important for workers in a noise exposure environment.In the subgroup analysis, we found that heterogeneity mainly came from cross-sectional studies. It is known that the results of cohort and case-control studies are more reliable. But there were no differences in results between the three types of studies. The subgroup analysis by gender showed that the positive association between smoking and NIHL was stronger in male than female. Because the smoking group of women was small, we only included one study focusing on female participants. And in general, men smoke more than women. Further, in the working environment, men generally have a higher noise exposure dose and a longer duration. Some studies suggested that men were more likely to suffer from NIHL as well ,52,53. AEgger’s linear regression test and Begg’s rank correlation test for the study of current smokers and NIHL were both suggestive of publication bias, which may be related to the inclusion of articles included published in Chinese and English only. In the process of paper screening, four studies published in other languages were eliminated. There was no search for unpublished grey documents. To address the potential of publication bias, we conducted the trim-and-fill method to adjust the influence of publication bias. There was no substantial change, suggesting that the result was not affected by this bias.Despite previous meta-analysis  suggestiThere are still some limitations in the present study. First, due to the lack of cohort and case-control studies, we added cross-sectional studies. However, cross-sectional studies have a selection bias due to the defects of its design, so that the results are not as reliable as the other two study designs. Secondly, the diagnostic criteria of NIHL were entirely different in each study. Some chose high-frequency (3 kHz) hearing loss as the standard, and some were based on speech frequency (−2 kHz) hearing loss. Moreover, the frequency selections of hearing thresholds were varied as well, which might lead to the heterogeneity. Third, there are also significant differences in the correction of factors that may affect the relationship between smoking and NIHL. The previous studies may not have adjusted all of the confounding factors similarly, such as heredity, with a great effect on NIHL. Fourth, in the dose-response analysis, few of included studies had high dose results, which affected the stability of the rear part of the curve.According to the shortcomings of our research, we hope to have more long-term follow-up prospective cohort studies to explore the relationship between smoking and NIHL to further confirm our conclusions. They also need uniform diagnostic criteria. In addition, the synergistic effect between smoking and noise interests us. We look forward to seeing a subgroup analysis of diverse occupational noise exposure history. Based on the available information, we have reason to believe that smoking is a risk factor for many diseases, and it can affect workers’ hearing health. Relevant departments can provide some smoking cessation programs in the occupational noise environment, especially for men. The Government can also reduce tobacco consumption by increasing tobacco taxes.In summary, our study indicated that smoking is a risk factor for NIHL. Quitting smoking can reduce the risk of NIHL. There is a non-linear dose-response relationship between the number of smoking pack-years and NIHL. When the dose is less than 15, the risk will add over the increase of pack-years.


The methylenetetrahydrofolate reductase (MTHFR) rs1801131 A/C variant results in a decrease in MTHFR enzymatic activity, which may play an important role in folate metabolism and is also an important source of DNA methylation and DNA synthesis. Several case-control studies have been conducted to assess the association of MTHFR rs1801131 polymorphism with the risk of urinary cancers, yet with conflicting conclusions. To derive a more precise estimation of above relationship, the association between the MTHFR rs1801131 A/C polymorphism and the risk of urinary cancer was performed.A total of 28 case-control studies was identified. The odds ratios (OR) with 95% confidence intervals (CI) was calculated to assess.On one hand, we found that the MTHFR rs1801131 A/C polymorphism was associated with increased whole urinary cancers’ risk (for example CA vs. AA: OR = 1.12. 95%CI = 1.01–1.24). On the other hand, we found that the MTHFR rs1801131 A/C polymorphism might increase bladder cancer risk both in Asian  and African populations (CA vs. AA: OR = 1.63. 95%CI = 1.17–2.25).Our current analysis suggested that MTHFR rs1801131 A/C is associated with urinary cancers, especially bladder cancer. Previous epidemiological studies have shown an association between low folate intake and an increased urinary cancer risk , 2, meanA common single nucleotide polymorphism (SNP), A1298C/rs1801131 A/C, is located in the coding carboxy-terminal regulatory region domain  and lympPrevious studies have investigated that MTHFR rs1801131 A/C was involved in the development of urinary cancers. However, the results of these studies remain conflicting. With the aim to measure the correlation, we performed this comprehensive meta-analysis by adopting all eligible studies –34.We searched the Pubmed database , using combinations of the keywords: ‘polymorphism,’ or ‘variant’ or ‘mutation’ and ‘bladder cancer’ or ‘prostate cancer’ or ‘renal’ and ‘MTHFR’ or ‘methylenetetrahydrofolate reductase’. All the included studies met the following criteria (1) the association between MTHFR rs1801131 A/C and urinary cancer risk was evaluated; (2) case-control studies were designed; (3) available genotype frequency was collected. The major exclusion criteria were (1) duplications; (2) insufficient reporting data; (3) abstract, commentary, review, editorial article and conference article.Two authors carefully extracted data from all eligible publications, independently. The following data were collected from each study: first author’s last name, year of publication, race of origin, cancer type, sample size (cases/controls), study design , source of control for prostate cancer subgroup, Hardy-Weinberg equilibrium (HWE) of controls and genotype method.The Newcastle-Ottawa Score (NOS) were selected to assess the quality of each study and to assess the various aspects of the methodology used by the observational research, which are relevant to the quality of the study, including the selection of cases, the comparability of groups and the determination of exposure. The total score is from 0 to 9 star. Studies with scores more than 7 are to be as high quality .Odd ratio (OR) with 95% confidence interval (CI) was used to measure the strength of the association between rs1801131 A/C and urinary cancers. Four different genetic models were applied to evaluate above association: allelic contrast , heterozygote comparison (CA vs. AA), dominant genetic model (CC + CA vs. AA), and recessive genetic model (CC vs. CA + AA). The ethnic descents were categorized as Caucasian, Asian, African, or Mixed. The control group based on sources was divided as follows: HB, PB, benign prostatic hyperplasia (BPH), and healthy man.Z-test. The heterogeneity was evaluated by both Cochrane Q-test [I2 metric [P for the heterogeneity test (Ph) < 0.10 and I2 > 50% [The statistical significance of the summary OR was determined with the e Q-test , 37 and 2 metric , 39 rangI2 > 50% , the pooI2 > 50% , 42.Subgroup analysis was performed according to the ethnicity and the source of cases to explore potential heterogeneity. The meta-regression analysis is a technique used to assess heterogeneity between the studies . This stThe PolyPhen-2 bioinformatic tool was used to predict the effects of gene SNPs on the translated proteins. In the PolyPhen-2 analysis, the scores could range from 0 to 1, where a score of zero meant ‘benign’ and a score of one meant ‘probably damaging’.http://string-db.org/) [The network of gene-gene interaction for MTHFR gene was utilized through String online server (db.org/) .https://www.ncbi.nlm.nih.gov/snp/rs1801131#frequency_tab): East Asian (EAS), 0.219; European (EUR), 0.313; African (AFR), 0.151; American (AMR), 0.15; and South Asian (SAS), 0.42  et al.  in our aP = 0.387 for heterogeneity, Fig. P = 0.003 for heterogeneity, Fig. In the total analysis, significant increased relationship was found in both heterozygote comparison  as dependent variable, the random-effect meta-regression results were presented in Fig. P = 0.119 for Begg’s test; t = 1.01, P = 0.314 for Egger’s test, Figs. Begg’s funnel plot and Egger’s test were performed to access the publication bias of the literature. The shape of the funnel plot did not reveal obvious asymmetry and the Egger’s test suggested the absence of publication bias [for example (CA vs. AA)  had a decreased association in whole cancer risk and urinary cancers , 48. AboIn addition, we used the online analysis system-String to predict potential and functional partners Fig. . FinallyThere are some limitations inherent in the included studies. First, despite inclusion of all the eligible studies, the resultant sample size is still not large enough; this situation may increase the likelihood of type I and type II errors. Second, we just searched articles from Pubmed, some other studies maybe omitted. Third, the cancer may not be depended entirely on a gene or one kind of polymorphism, because different results were found in rs1801131 polymorphism, and in different SNPs (such as rs1801133 polymorphism) in the same MTHFR gene in current analysis, further studies should be to identified more valuable and credible polymorphisms. Fourth, it is necessary to evaluate the roles of some special environmental factors  and lifestyles. Fifth, significant associations were detected in different genetic models in the same subgroup, this inconsistency may indicate the influence of type I error by the repetitive comparison.In summary, our present update analysis suggested novel evidence that the MTHFR rs1801131 polymorphism has a risk effect for urinary cancers, especially bladder cancer. Further studies with larger samples, are needed to evaluate associations between MTHFR rs1801131 polymorphism and urinary cancers’ risk.


In the meta-ANOVA and -regression test, administering chitosan and resistant starch was revealed to be most effective in lowering body weight. In addition, chitosan supplementation proved to be an effective solution for serum TNF-α inhibition. In conclusion, chitosan has been shown to be somewhat useful in improving symptoms of lifestyle-related disease. Although there are some limitations in the results of this meta-analysis due to the limited number of animal experiments conducted, chitosan administration nevertheless shows promise in reducing the risk of cholesterol related metabolic disorder.This study presents a meta-analysis of studies that investigate the effectiveness of chitosan administration on lifestyle-related disease in murine models. A total of 34 published studies were used to evaluate the effect of chitosan supplementation. The effect sizes for various items after chitosan administration were evaluated using the standardized mean difference. Using Cochran’s Q test, the heterogeneity of effect sizes was assessed, after which a meta-ANOVA and -regression test was conducted to explain the heterogeneity of effect sizes using the mixed-effect model. Publication bias was performed using Egger’s linear regression test. Among the items evaluated, blood triglyceride and HDL-cholesterol showed the highest heterogeneity, respectively. Other than blood HDL-cholesterol, total cholesterol, and triglyceride in feces, most items evaluated showed a negative effect size with high significance in the fixed- and random-effect model ( Lifestyle-related diseases, including obesity, hyperlipidemia, atherosclerosis, type II diabetes, and hypertension, are widespread in industrialized countries, and are major threats to cardiovascular health. The syndrome is related to a combination of metabolic disorders, including abdominal obesity, hypertriglyceridemia, high-density lipoprotein (HDL) cholesterol decrease, hypertension, and high blood glucose, which lead to increased cardiovascular morbidity and mortality . UnnaturChitosan is one of the polymers containing acetyl glucosamine and glucosamine. It may be obtained by hydrolyzing and converting chitin with alkali from crabs, shrimps, insects, mushrooms, and the cell walls of microorganisms. Chitosan manufacture by deacetylation of chitin has been utilized in wastewater treatment and the agricultural sector. As the safety of chitin or chitosan has become increasingly recognized, it has recently-been used in a variety of fields, including medical supplies, food additives, and cosmetics ,4. ChitoMeta-analysis is a method of statistical analysis that combines results from various scientific studies to obtain a quantified synthesis . Meta-anStudies of lifestyle diseases in murine models suggest that they may be improved by administering chitosan. However, few comprehensive studies have been conducted to date on the effect of chitosan supplementation on improving lifestyle diseases. Accordingly, the objective of the present study was to perform a meta-analysis of the effects of chitosan on factors in lifestyle-related diseases in adults.p < 0.0001) in both effect models. These results mean that chitosan administration results in decreased levels of TC, TG, and LDL-C in blood, TC and TG in the liver, serum TNF-α and glucose in blood and body weight, and increased levels of blood HDL-C, fecal TC and TG.The effect sizes of chitosan administration on hyperlipidemia in murine models using fixed and random effect models are listed in b) calculated under assessing that between subgroups (τ2) was the same. First of all, a meta-ANOVA analysis was performed on most items except fecal TG, as shown in p < 0.05). p > 0.05). Other than fecal TC, body weight, and blood glucose, most items showed significant differences (p < 0.05). In the case of TC, the Q statistics between the groups (Qb) was 31.94 ; the effect size between groups was assumed to be significantly different.Since heterogeneity analysis in this study revealed a high level of heterogeneity between the studies analyzed, moderator analysis was performed to account for this. For this, meta-ANOVA and meta-regression were conducted. To perform the meta-ANOVA test, Q statistics between the subgroups (Qp = 0.0208), and it was revealed to use to decrease blood TC. In the case of WSC, it was significantly effective in lowering of serum TNF-α and body weight . With regard to the administration period, this was significantly relevant to blood HDL-C and liver TC with p = 0.0004 and 0.0358, respectively.Next, meta-regression was performed to evaluate the effect size between the type of chitosan used and the administration period . Only CTp < 0.05) indicating that the relationship between effect size and standard error was statistically significant and confirming the presence of bias [p < 0.05).Publication bias was conducted using an Egger’s linear regression test  on blood of bias . Thus, tp < 0.0001) in both effect models. These results mean that chitosan administration results in decreased levels of TC, TG, and LDL-C in blood, TC and TG in the liver, TNF-α and glucose in blood and body weight, and increased levels of blood HDL-C, fecal TC and TG.In the results of 3+), its viscosity is increased by the formation of poly cations and gels. In high viscosity of the intestine, dietary fiber lower blood cholesterol by delaying cholesterol diffusion from micelle to mucosa, inhibiting bile acid metabolism, delaying micelle forming, and reducing cholesterol absorption rate in the intestine [The bioavailability of dietary fat in the intestine decreased after chitosan administration. After this, reverse cholesterol transport, which is delivered from peripheral tissues to the liver, is accelerated by excretion of surplus dietary fat, resulting in an increase in the ratio of HDL-cholesterol . Similarntestine ,53. BaseCytokines are secreted by activated lymphocytes and macrophages, and regulate the function of the cells related to immune response. They are also recognized as playing an essential role in the inflammatory response . Yemak eAccording to Prabu and Naturajan , blood g2 values were computed. The highest among Q statistics was TG in blood, with high significance (p < 0.0001). The significance of the Q statistic implies that the studies used to calculate the overall effect (the effect size of fixed and random effect models) do not share the same effect size with one another . However, one limitation of this method is its dependence on the number of studies . In thi (Fleiss ). I2 ands et al. ). All itogeneity . The τ2 ct sizes . In addiCholestyramine  and cholestipol (trade name: Colestid or Cholestabyl) as an anion-exchanger are these days used mainly for reducing cholesterol . These mTo perform a meta-analysis of published studies regarding the effect of chitosan administration on lowering cholesterol in murine models between 1978 to 2020, a literature search was conducted on Pubmed  and Science Direct . The keywords used for searching studies for meta-analysis were “chitosan, cholesterol” in all databases. The results obtained included 450 citations from Science Direct and 303 from Pubmed . These results were then filtered by title, abstract, and full text. Among them, 4 review articles and 7 studies of clinical tests in human studies were removed. Also, the studies expressed with graphical data were eliminated. Following this, studies regarding changes in cholesterol levels after chitosan administration were collected. Ultimately, a total of 34 studies with 11 items  were selected to perform a meta-analysis of the effectiveness of chitosan in reducing cholesterol in murine models.2 value. To confirm the heterogeneity of effect size using a mixed-effect model for the items in question, meta-ANOVA and regression analyses were also used. Meta-ANOVA and meta-regression analysis can evaluate the difference of Hedges’ g among subgroups herein administration periods or type of treatment. The periods were set as independent factors in meta-ANOVA and as continuous variables in meta-regression. Finally, publication bias analysis was conducted to ensure the validity of the meta-analysis results. Statistical analysis and visualization of the results were performed using the ‘meta’, and ‘metafor’ packages in the R statistics software application .Corrected standardized mean difference (Hedges’ g), and 95% confidence intervals (CI) were computed between control groups and treatment. The weight of the effect size was calculated using inverse-variance ,81. EffeThe present study confirmed the effectiveness of chitosan administration on lifestyle-related diseases through meta-analysis. Chitosan was significantly effective in lowering total cholesterol and triglyceride of blood and liver and rising fecal total cholesterol and triglyceride. Based on our results, chitosan was demonstrated to be useful in improving the symptoms of lifestyle-related disease.


Diets are complex to examine as they are multi-dimensional and often accompanied by internal correlations. The use of dietary indices to quantify dietary quality and/or variance via a single, simplified measure has gained traction in the last two decades. Several diet indices have been developed to evaluate the role of diet in relation to chronic disease mortality, with most originally based on data from US and Europe populations. Following which, a number of studies have emerged to investigate the compatibility and applicability of various indices for different health outcomes and for different population groups. Current evidence suggests that certain indices maybe more suitable for examining certain health outcomes or populations, as different indices explain different variance in food intakes and/or that certain dietary components captured may not be as relevant to the investigated health outcomes or cultural context , 2.The study by Chen L-W et al. brings to mind this pertinent point, but also brings to the table a refreshing perspective of the value of using different dietary indices to understand the various mechanisms diets may act through, in relation to health . By asseThe dietary index of particular interest here is the DII, which has been applied to various clinical conditions, but only more recently been extended to the area of maternal and child health. The DII was developed and validated to characterize and quantify the cumulative inflammatory potential of individual diet. One of its key advantages is that it measures the standardized contribution of a wide variety of foods and nutrients based on inflammatory biomarkers objectively measured in cell, animal, or human studies. This allows it to be universally applied across many cultural contexts and derived from different dietary assessment tools, thus allowing comparison of studies globally .n = 16,295 mother-child pairs—derived from using individual participant data pooled from seven European cohorts in the ALPHABET consortium. As opposed to traditional meta-analysis, the commendable effort in harmonizing data across multiple studies would have reduced substantial heterogeneity across studies due to incomplete or selective reporting and consequently facilitated more accurate interpretation of results. Adequately powered sensitivity analyses were also conducted, thus increasing the robustness of the findings.While the relationship of maternal or early life pro-inflammatory diets (measured using the DII) and child adiposity has previously been examined in two other cohort studies with approximately 1000 mother-child pairs each , 6, a reMeta-analyses are an integral part of evidence-based health research and whilst it is recognized that individual participant data meta-analysis offers numerous clinical and statistical advantages, one should be cognizant of its multiple challenges of being resource-intensive, the need for advanced statistical expertise and intensive co-ordination across groups/studies, amongst other challenges, before embarking on one . Replica


Objective: Epidemiologic studies investigating the association between dietary carbohydrates as well as glycemic index and glycemic load  and bladder cancer risk have yielded inconsistent results. The aim of the present meta-analysis is to summarize the evidence on this association.Materials and Methods: A comprehensive literature search of articles published by December 2019 was performed in PubMed, Scopus, and Web of Science databases. A random-effects model was used to calculate the pooled odds ratios (ORs) and their corresponding 95% confidence intervals (CIs).Results: Twelve observational studies were included in the final analysis. There was no evidence of an association between consumption of carbohydrates and bladder cancer risk . No statistically significant association between glycemic load and bladder cancer was likewise found . However, there was a significant positive association between glycemic index and bladder cancer risk . In the dose–response analysis, the pooled OR (95% CI) per 10 units of glycemic index per day was 1.02 .Conclusion: In this meta-analysis, glycemic index showed a positive linear association with bladder cancer risk. Bladder cancer is a common human malignancy, with an estimated 429,800 new cancer cases and 165,100 deaths in 2012 worldwide . EmerginCarbohydrates are one of the main dietary components affecting an individual's insulin secretion and blood glycemic response . GlycemiA comprehensive literature search of articles published by December 2019 was performed in PubMed, Scopus, and Web of Science databases with the following search algorithm: (diet or dietary or nutrition or nutrient or food or glycemic index or glycemic load or glycemic index or glycemic load or carbohydrate or carbohydrates) and  and  and (cohort or case–cohort or case–control or case control). We also examined the reference lists from retrieved articles and other relevant reviews to identify any additional studies. This study was performed according to the Preferred Reporting Items for Systematic Reviews and Meta-Analyses statement .The inclusion criteria were as follows: (1) investigation of the association between dietary carbohydrates, glycemic index, or glycemic load and bladder cancer risk, (2) use of the following study design: case–control, nested case–control, cohort, and case–cohort studies as well as follow-up studies of randomized clinical trials, and (3) reporting of fully adjusted risk estimates, including relative risk (RR), hazard ratio (HR), or odds ratio (OR), and the corresponding 95% confidence intervals (CIs) for the association between bladder cancer and dietary carbohydrate, glycemic index, or glycemic load. If multiple publications used the same study population, the latest publication providing the largest sample size was included. The Participants, Intervention Comparators, Outcomes, Study Design criteria are presented in The following data were extracted by two independent investigators (HZ and XX): first author's surname, publication year, country where the study was performed, specific characteristics of the study population, study design, age of participants, study size, number of cases, duration of follow-up for prospective studies, dietary assessment method, exposure , quantity of intake, risk estimates, 95% CIs from the most fully adjusted models, and matched or adjusted variables .http://www.ohri.ca/programs/clinical_epidemiology/oxford.asp). Any disagreements were resolved by discussing with a third reviewer. NOS is an eight-item instrument that allows for the assessment of the study population and selection, study comparability, follow-up, and the outcome of interest. The range of possible scores is 0 to 9. Scores <7 and ≥7 were assigned for low- and high-quality studies, respectively.The quality of each study was assessed by the same two investigators (HZ and XX) using the Newcastle–Ottawa Scale (NOS)  and the risk of bladder cancer were examined by using a random-effects model . The posThe linear dose–response trends were computed from the natural logarithm of the ORs and 95% CIs across the categories of exposure (at least three categories) using the method proposed by Greenland and Longnecker . In addiQ and I2 statistics, with the significant level set at P ≤ 0.1 .Heterogeneity between studies was explored using  P ≤ 0.1 . Publica P ≤ 0.1  and Egge P ≤ 0.1 . A two-tThe process of literature search and selection is presented in I2 = 0.0% and Pheterogeneity = 0.618). In the subgroup analyses, neither geographic region nor study design modified the association between carbohydrate intake and bladder cancer risk . There was no evidence of differences by geographic area and study design . No difference was observed when data were stratified by geographic area and study design (Four studies –14 were y design . Leave-oy design .Pnonlinearity = 0.389). The curve showed a significant increase in bladder cancer risk with increasing units of glycemic index. The pooled OR (95% CI) per 10 units of glycemic index per day was 1.02  (We then further explored if there was a dose–response relation between dietary glycemic index and bladder cancer risk. There was no evidence of a non-linear association between glycemic index and bladder cancer risk (01–1.04) .In this meta-analysis of observational studies, we found that a higher glycemic index was significantly related to a greater risk of bladder cancer. No association was observed between risk of bladder cancer and carbohydrate intake or glycemic load. To the best of our knowledge, this is the first meta-analysis to evaluate the association between dietary carbohydrates, glycemic index, or glycemic load and bladder cancer risk.Recently, the relationship between carbohydrates, glycemic index, or glycemic load and the risk of human cancers has been widely studied. Schlesinger et al.  and MullOur study indicated that there was a positive linear association between glycemic index and bladder cancer risk. Glycemic index is an indicator of the quality of carbohydrate foods. Foods with high glycemic index are rapidly digested and lead to a rapid peak in blood glucose and the secretion of insulin . HyperinThis meta-analysis has several strengths. First, both high- and low-level analyses and dose–response analysis were performed for the association between glycemic index and bladder cancer risk, with consistent results. Second, there was no evidence of obvious heterogeneity across studies. Third, we performed a throughout literature search, and no publication bias was detected. Fourth, leave-one-out sensitivity analyses showed no marked difference in results, indicating that the pooled risk estimates were robust and not driven by any single study.As with any study, this meta-analysis also has some limitations. First, individuals with a diet of high carbohydrate intake, glycemic index, or glycemic load may also have other unhealthy behavioral and dietary habits and metabolic-related illness, such as lack of physical activity, excess intake of total energy, diabetes, and obesity, which may confound the true association. Second, measurement error of diet may exist. Food frequency questionnaires are not originally designed to assess glycemic index and glycemic load. Most dietary questionnaires have calculated the glycemic index or glycemic load values according to a limited number of food items, which may affect the accuracy of exposure assessment. Finally, the inclusion of case–control studies may lead to recall and selection bias.In conclusion, in this meta-analysis, glycemic index showed a positive linear association with bladder cancer risk. Carbohydrate intake and glycemic load were not associated with the risk of bladder cancer. Furthermore, large prospective studies that account for other relevant factors, such as body weight, diabetes, physical activity, and intake of total energy, are needed.The datasets generated for this study are available on request to the corresponding author.XX and BL contributed to the conception or design of the work. HZ, QM, and HS contributed to the acquisition, analysis, or interpretation of data for the work. XX and SW drafted the manuscript. HZ and BL critically revised the manuscript. All authors gave final approval and agree to be accountable for all aspects of work ensuring integrity and accuracy.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 handling editor declared a shared affiliation with the authors at time of review.


MDR1 gene plays a crucial role in the absorption, transport, metabolism and elimination of harmful compounds. An impaired metabolism of these compounds related to genetic polymorphism may cause cancer such as acute myeloid leukemia AML. The human multidrug resistance C1236T polymorphism and the risk of AML development in a sample of Moroccan population. The present study investigated the relationship between MDR1 C1236T polymorphism was identified by PCR-RFLP method. Meta-analysis was performed to discuss our results. Statistical analyses were performed using SPSS, MetaGenyo and MedCalc. The present case-control study included 131 AML patients and 136 healthy controls. The 1236TT mutant genotype and the risk of AML  compared to the wild type 1236CC. In addition, the recessive model revealed that carriers of 1236TT mutant genotype were more exposed to develop AML when compared to the combined 1236CC/CT genotype . The clinical parameters of AML showed no significant association. Meta-analysis demonstrated no statistically significant association between this polymorphism and AML susceptibility. A positive association was found between the MDR1C1236T polymorphism appears to be associated with the risk of AML. Further studies, including a large sample size, are needed to confirm these findings.Our study suggests that the  Acute myeloid leukemia (AML) is the most deadly type of leukemia . As other types of cancer, the etiology of AML remains unknown. However, many risk factors were incriminated such as exposition to pesticides, radiotherapy, chemotherapy, benzene and radiation . In addition, many genes have been reported to be associated with higher risk of AML.MDR1 or ABCB1 gene is expressed in organs implicated in xenobiotics metabolism and excretion . Therefore, MDR1 plays a crucial role in the elimination of carcinogenic substances. The alteration of MDR1 activity related to interindividual genetics background may lead to the development of cancers .MDR1 gene is located on chromosome 7, at position q21, which encodes for a transmembrane protein that plays role of ATP -dependent efflux transporter pump, called P-glycoprotein (P-gp) . MDR1 is a highly polymorphic gene. About a thousand single nucleotide polymorphisms (SNPs) have been found .In humans, the SNP C1236T in exon12 (rs1128503) is one of the most studied SNPs in this gene, that leads to the alteration of MDR1 expression and activity . The allele frequencies of this polymorphism vary from 30% to 93% between different ethnic groups worldwide  Several genetic epidemiological studies reported a positive association between this polymorphism in MDR1 gene and cancer risk such us, acute lymphoblastic leukemia, breast cancer, acute myeloid leukemia and non-hodgkin lymphoma The synonymous C1236T polymorphism and AML susceptibility.The purpose of the present study was to explore the relationship between Patients and MethodsStudy populationThe present case–control study included 131 AML patients and 136 healthy age-matched unrelated controls, recruited from the Department of Hematology and Pediatric Oncology, 20 August Hospital, University Hospital, Ibn Rochd, Casablanca Morocco from 2015 to 2017. Individual written informed consent was obtained from all participants. The classification of AML subtypes in patients was performed according to the criteria of the World Health Organization 2008 (WHO) .Genotyping of C1236T C1236T MDR1 polymorphism was performed using polymerase chain reaction followed by the restriction fragment length polymorphism (PCR-RFLP) technique previously described by Kassogue et al., (2014). The PCR reaction was carried out in a final volume of 25µl. The PCR products were digested overnight with HaeIII and analyzed on a 3% agarose gel stained with ethidium bromide.Genomic DNA was extracted by the salting-out protocol . DNA concentration was estimated by NanoVue Plus spectrophotometer. Genotypic identification of the Meta-AnalysisRetrieval strategyMDR1, ABCB1, acute myeloid leukemia, AML, polymorphism, C1236T, SNP. Eligible studies were found by searching PubMed, Science Direct and Google Scholar. Literature search was accomplished independently by two investigators to find all articles published before March 27, 2019 that investigating the relationship between the polymorphisms and AML risk, using the following key words: P-gp, multidrug resistance gene, Inclusion and exclusion criteriaMDR1 C1236T polymorphism and AML with sufficient data were included. The exclusion criteria were: chronic myeloid leukemia, acute lymphoid leukemia  and lymphoma studies, review articles, studies that did not meet our study and duplicate studies.The inclusion criteria were: studies about the association between the Pooled Studies for the Meta-Analysis MDR1 C1236T polymorphism in AML patients were included in the present meta-analysis with 480 cases and 920 controls Four case-control studies investigating Statistical analysisC1236T MDR1 polymorphism and AML risk. We used either the chi-square or Fisher’s exact tests to evaluate the association between C1236T MDR1 polymorphism and clinical parameters; a p value less than 0.05 was considered statistically significant. The dominant genetic model was defined as TT homozygote + CT heterozygote mutant genotypes compared with homozygote Wild-type genotype CC, while the recessive genetic model was defined as TT homozygote mutant compared with heterozygote+ homozygote Wild-type genotypes. For the meta-analysis the ORs were estimated using a fixed-effects model. Genetic Heterogeneity was estimated with the Cochran’s Q test and I2 statistics was used to estimate the heterogeneity among studies. A random-effects model was used to estimate common OR if heterogeneity was detected; I2> 50% indicated heterogeneity between studies.Statistical analyzes were performed using SPSS  and MedCalc  as well as MetaGeny (GENYO). The Hardy–Weinberg equilibrium was performed by chi-square test in patients and controls. The odds ratio (OR) with 95% confidence intervals (CI) was calculated to assess the association between the C1236T MDR1 gene polymorphism on the risk of developing AML in 131 patients with AML and 136 healthy controls. The genotypic distribution of C1236T MDR1 gene polymorphism did not deviate from Hardy-Weinberg Equilibrium in controls and patients . MDR1 C1236T, we observed no statistically significant association between this polymorphism and AML under different genetic models (P > 0.05). No heterogeneity between studies was found, I2 =0% Based on the studies published on MDR1 C1236T polymorphism on acute myeloid leukemia risk in a sample of Moroccan population and we found that the mutant TT genotype of MDR1 C1236T polymorphism was associated with the susceptibility of acute myeloid leukemia when compared to the wild type CC genotype. The most frequently observed FAB subtype in our study was M2; this result is consistent with our previous studies . Other studies have reported similar results . Regarding the correlation between this polymorphism and clinical parameters, we found that there was no statistically significant positive association between the MDR1 C1236T polymorphism and gender, age, karyotype and FAB types. Interestingly, our results showed a significant association between the TT homozygote mutant genotype and the risk of AML with an odds ratio of 2.39 . AL-Faisal et al. reported a similar result in Iraqi population .In this case-control study, we investigated the impact of MDR1 C1236T gene polymorphism and AML risk. Feng et al., (2016) reported that no significant relationship between the ABCB1 C1236T variant and AML. Green et al., (2012) also found no significant difference in genotype frequencies between AML patients and the reference population. The difference in results between studies of MDR1 C1236T gene polymorphism could be explained by many factors such as the genetic background, the small sample size, difference in genotyping methods, and the mode of living and it should be noted that the frequency of this SNP varied according the ethnic group and to geographic location .In contrast, our Meta-Analysis and some studies have not shown an influence of this polymorphism and the risk of developing AML. Jingdong et al., (2016) reported that there is no significant association between MDR1 C1236T gene polymorphism on the risk of AML could be explained by the intracellular accumulation of toxic and carcinogenic substances due to the low expression of MDR1 gene. Furthermore, many studies reported that the diminished MDR1 expression leads to increase DNA damage and Cytogenetic aberrations that play a critical role in the leukemogenesis .The influence of MDR1 C1236T polymorphism could influence the risk of developing adult acute myeloid leukemia. Future studies with large sample size are necessary to confirm the association between MDR1 C1236T polymorphism and AML risk. In conclusion, this study results suggested that the The present case-control study was approved by the Ethical Committee of Hassan II University, School of Medicine and Pharmacy, Casablanca, Morocco, N°02/19. A written informed consent was obtained from all participants before entering the study


This systematic review and meta-analysis aimed at providing quantitative evidence for the effects of BCX on osteoporosis. Publications were selected and retrieved from three databases and carefully screened to evaluate their eligibility. Data from the final 15 eligible studies were extracted and uniformly summarized. Among the 15 studies, seven including 100,496 individuals provided information for the meta-analysis. A random effects model was applied to integrate the odds ratio (OR) to compare the risk of osteoporosis and osteoporosis-related complications between the groups with high and low intake of BCX. A high intake of BCX was significantly correlated with a reduced risk of osteoporosis  0.70–0.90, p = 0.0002). The results remained significant when patients were stratified into male and female subgroups as well as Western and Asian cohorts. A high intake of BCX was also negatively associated with the incidence of hip fracture . The results indicate that BCX intake potentially reduces the risk of osteoporosis and hip fracture. Further longitudinal studies are needed to validate the causality of current findings.Many studies have analyzed the effects of  Osteoporosis is a chronic and systematic skeletal disease characterized by reduced bone mineral density (BMD), eventually leading to bone fragility and bone fracture . Hip fraβ-carotene. Similar to other carotenoids, BCX is susceptible to oxidative and enzymatic cleavage, which subsequently forms many bioactive metabolites ..19].β-cryptoxanthin” or “beta-cryptoxanthin” or “cryptoxanthin” or “cryptoxanthine” or “beta-caroten-3-ol” or “carotene-3-ol” or “carotenol”) and (“osteoporosis” or “osteoporotic” OR “osteopenia” or “osteoblast” or “osteoclast” or “bone”). The search was conducted on the first day of July and was not limited to their publication dates. All types of clinical observational studies investigating BCX with regard to osteoporotic-related health problems were defined as eligible. After being imported into Endnote X9, duplicates were screened and filtered. The remaining studies were screened for their titles, abstracts, and keywords to examine their eligibility. A study was excluded for the following reasons: (1) the inclusion criteria were violated, (2) only irrelevant data was presented to extract, (3) the topic was unrelated (4) it was not a peer-reviewed research article, (5) it was duplicated entirely in both cohort and subject with another study, or (6) the full text was unavailable. Manual screening on the reference list of previous articles was conducted, and periodic checks were carried out after the search date to complement additional studies.Studies were collected by searching on international literature databases including PubMed, Embase, and Cochrane Central Register of Controlled Trials (CENTRAL) with the following search terms: , and sample size . Regarding outcome measurement, data of evaluation parameters, statistical models, adjusted factors for statistical analysis, and primary outcomes were extracted.We primarily conducted a meta-analysis to evaluate the overall association between osteoporosis-related outcomes and BCX intake. We included only studies that estimated the association of osteoporotic risk-related outcomes reported as odds ratio (OR) and 95% confidence interval (95% CI). Studies comparing bone-related risk between the highest-level and the lowest-level groups  were included in the pooled analyses. When multiple correlation models were reported from a single cohort, the adjusted model that encompassed the most confounding variables was used. If a study did not provide an adjusted OR, imputation was applied by calculating the raw OR using the provided data. Owing to the lack of complete information from the study of Sahni et al., ORs from the previously published meta-analysis were employed in the current data synthesis .I2 statistic. A funnel plot was not applied in this study as well as other analyses of publication bias due to the small number of included studies.The data were synthesized using RevMan 5 (the Cochrane Collaboration). ORs from each publication were pooled using the generic inverse variance method with the random-effects analytic model. Heterogeneity across the studies was reported using the Cochran Q test and Leave-one-out sensitivity analysis was conducted to inspect the robustness of the metadata. Sensitivity analysis was performed for all the outcomes of the meta-analysis, including pooled ORs of the overall osteoporotic risk and its subgroups.Methodological quality was evaluated based on the study types using the tools from the National Health Institute (NIH). Critical study design-specific aspects related to the risk of bias were independently inspected for the included studies. Each item was scored as “Yes” or “No” according to the qualification. Otherwise, they were classified as “cannot determine (CD)”, “not applicable (NA)”, or “not reported (NR)” depending on the information provided. After evaluation, each author provided an overall quality rating for each study. The conflict was solved by the third reviewer’s arbitration.The overall workflow of the current systematic review and meta-analysis is presented in The methodological qualities of all studies were rated as “fair” to “good” . HoweverThe detailed demographic characteristics, evaluation parameters, and outcomes are described in p = 0.0002) (p = 0.11). Due to the physiological differences between males and females, subgroup analysis stratified by sex was conducted. As a result, we observed that the high intake of BCX in female subjects was significantly correlated with a lower risk of osteoporosis than that in the lower consumption group . A similar result was observed in male subjects but to a smaller extent of reduced osteoporotic risk . The heterogeneity in the male subgroup population was relatively low (I2 = 11%). In addition, there was no significant heterogeneity between the male and female subgroups (p = 0.79).Seven studies including 100,496 participants were included in the meta-analysis ,22,23,24 0.0002) . The ovep = 0.02, I2 = 0%)   a. Asian 2 = 56%) b.p = 0.0008, I2 = 55%) . A higher intake of BCX was also significantly correlated with a lower risk of hip fracture in the female subgroup . No significant difference was found between the groups (p = 0.99).Because hip fracture is one of the most common complications of osteoporosis, we additionally conducted a meta-analysis focusing on this particular fracture position. In total, four studies recruiting 76,663 participants provided sufficient information for the meta-analysis. All included studies explored the outcome of both sexes, with ages ranging from 39 to 80 years. Therefore, a subgroup analysis based on sex was also conducted. Overall, the results revealed that a higher BCX intake was significantly correlated with fewer events of hip fracture  . ConsideMale and Western ethnic subgroups, based on the overall risk of osteoporosis, were susceptible to changes according to the leave-one-out sensitivity analysis. Meanwhile, both male and female subgroups were also sensitively affected in terms of the outcome of the risk of hip fracture. In all cases, elimination of the data from Cao et al. remarkably reduced the heterogeneity..We recorded eight other studies that reported an association between BCX and the incidence of osteoporosis. However, no data were available or suitable for conducting the meta-analysis. Of note, all eight studies investigated the association between blood BCX and osteoporotic risk. A summary of characteristics and key findings is given in Table 1Previous studies have suggested notable beneficial functions of BCX. Among them, the specific benefits of the effects of BCX on bone homeostasis have received more attention in recent years, particularly in clinical studies. Our primary result of the meta-analysis was that a high amount of dietary BCX was significantly correlated with a low risk of osteoporosis. In addition, corresponding outcomes were also drawn from the subgroup analyses. Females benefited more than males when they consumed a high amount of BCX. Another interesting finding was that a consistent and significant relationship was observed in different cohort ethnicity. However, due to the limited number of studies, the confidence and robustness of the results need to be further validated with quality control procedures by sensitivity analysis.Numerous studies have investigated natural products and compounds to discover their efficacy on osteoporosis ,34,35. IThe physiological aspects of osteoporotic risk depend on sex, particularly in postmenopausal women. Therefore, a daily diet with sufficient vitamins, calcium, and carotenoids is particularly crucial to postmenopausal women . Our finHip fracture, the most common osteoporosis-related complication, is a significant health problem leading to a decrease in life expectancy, disability, and mortality in the elderly population . In thisIn addition to the investigation of BCX intake, the correlation of serum BCX levels was also commonly explored. Considering that Sugiura et al. performed statistical analysis on the same cohort, four cohort populations were analyzed to investigate the correlation between BMD and BCX intake or serum BCX level ,26,27,30This study is the first systematic review and meta-analysis of the association between BCX intake and the overall risk of osteoporosis. Sex and ethnicity were also associated with the degree of risk reduction. Meanwhile, the result is an updated meta-analysis on the risk of hip fracture. There are several limitations to this systematic review and meta-analysis, which should be specified in the current investigation. First, the number of quantitatively synthesized studies was small, as only seven out of 15 studies were incorporated into the meta-analysis because of the impractical format of data, such as the beta coefficient of the regression model, overlapping cohort, or lack of relevant data. Second, imputation of the unavailable data might introduce an unexpected error in the data synthesis procedure. Third, another potential source of bias was the heterogeneity of study design across the included studies, as well as the comparison by combining either different fractiles of BCX intake or differently adjusted covariates. Likewise, outcome variables could not be matched due to the inconsistency in outcome measurements. The estimation of BCX intake, in the included studies, was difficult to clarify because of limited access to individual food questionnaires. Finally, multiple cross-sectional and case-control studies were involved, but the causality of the exposure and the event was not reported. Taken collectively, the current meta-analysis supports the previous idea regarding the potential benefit of BCX intake in lowering the risk of osteoporosis and fracture. Further investigations with a solid study design and consistent outcomes should be conducted.In conclusion, the evidence of this meta-analysis suggested that the intake of BCX was significantly associated with a lower risk of osteoporosis. Consistent trends were observed in the male and female subgroups and both Western and Asian cohorts. The risk of hip fracture was also negatively associated with BCX intake. Moreover, 15 observational studies that investigated the effect of BCX on osteoporotic risk were summarized with a detailed description of study characteristics. Collectively, our findings can guide the development of the functional use of BCX. Further longitudinal studies and randomized controlled trials on this aspect are required to address a higher level of evidence.