ACE) gene is indispensable for endothelial control and vascular tone regulatory systems, usually affected in Systemic Lupus Erythematosus (SLE). ACE insertion/deletion (I/D) polymorphism may influence the progress of SLE. Earlier studies have investigated this association without any consistency in results. We performed this meta-analysis to evaluate the precise association between ACE I/D polymorphism and SLE susceptibility. The relevant studies were searched until December, 2017 using Medline (PubMed), Google-Scholar and EMBASE search engines. Twenty-five published studies involving 3,308 cases and 4,235 controls were included in this meta-analysis. Statistically significant increased risk was found for allelic , homozygous , dominant , and recessive  genetic models. Subgroup analysis stratified by Asian ethnicity revealed significant risk of SLE in allelic  and marginal risk in dominant  models; whereas, no association was observed for Caucasian and African population. Publication bias was absent. In conclusion, ACE I/D polymorphism has significant role in overall SLE risk and it can be exploited as a prognostic marker for early SLE predisposition.Angiotensin-converting enzyme ( Systemic lupus erythematosus (SLE) is an autoimmune disease often involving multiple organ inflammation. The clinical consequences of SLE are extremely heterogeneous and generally characterized by pathogenic autoantibody formation against the host's nuclear antigens, immune complex deposition, and end-organ damage  or deletion (D) of a 287-bp fragment on intron 16  is a zinc metalloproteinase converting angiotensin I (Ang I) into angiotensin II (Ang II), an octapeptide acting as a potent vasopressor and stimulator of aldosterone. Ang II also acts as a growth factor, particularly in kidneys, inducing remodeling of tissue, and fibrosis. Contraction of smooth muscle cells and their proliferation is also known to be induced by Ang II, including adhesion of monocytes and platelets  for their quality score. Trial Sequential Analysis (TSA) was used to minimize type-I statistical errors like publication bias and random errors, caused generally by sparse data, in order to quantify the statistical reliability of the included data in the meta-analysis with statistical significance threshold.A meta-analysis, as a statistical tool, can overcome these limitations of single studies with small sample sizes. It combines multiple studies on the same alleles of genes to enhance the statistical power of the analysis and derive more precise and reliable results of the genetic effects. Therefore, we performed this meta-analysis by pooling all the eligible published studies to determine a comprehensive picture of the above said genetic association and understand the role of ACE I/D gene polymorphism and its association with SLE susceptibility was conducted in PubMed (Medline), Google Scholar and EMBASE. An update limit of December 2017 was applied. The search terms were as follows: “ACE” “angiotensin converting enzyme” OR “ACE” (polymorphism OR mutation OR variant) in combination with “systemic lupus erythematosus” OR “SLE” (susceptibility OR risk). The focus of the search was human studies only. The titles and abstracts of the all the searched articles were read for initial evaluation, and only studies fulfilling the eligibility criteria were retrieved and used in this meta-analysis. The reference lists of the selected studies was also inspected manually for other pertinent articles.A thorough search of studies on the ACE I/D gene polymorphism and SLE susceptibility, (b) the study must have a case-control design, (c) the study must have enrolled well diagnosed SLE patients and normal healthy controls, (d) the study must have genotypic frequency data available for both patients and controls, (e) the study must be published in English. In case the data for same patient population was reported in more than one publication, the most recent and complete publication was considered for this meta-analysis. The exclusion criteria for the studies were: studies with overlapping data; studies reporting data for patient population only; studies where genotypic frequency data was not available; and review articles. All information regarding selection of the studies is depicted as PRISMA 2009 Flow-Diagram  the study must be an evaluation of the association between m Figure .Two investigators (RKM & SAD), independently, extracted and summarized the data from each retrieved study by following the standard procedure. The data collection form was designed and used to collect the data to ensure accuracy while following stringent inclusion/exclusion criteria as described above. The attributes extracted and summarized from the selected articles were: first author's name; publication year; origin country; total number of patients and controls included; study type; association status; genotyping method used; and genotype frequency of patients and controls. Discrepancy, if observed, in the data collected from the selected studies by the two investigators was settled by open discussion in presence of SH (an adjudicator) in order to reach a final consensus.The NOS criteria Stang,  was usedp < 0.05 was considered as statistically significant. All the p-values were two sided. Crude odds ratios (ORs) and corresponding 95% confidence intervals (CIs) were computed to assess the association intensity between the ACE I/D gene polymorphism and SLE susceptibility. The allele contrast, log-additive, dominant, and recessive model pooled ORs were estimated  V2 software program . The d Woolf, . Chi-squd Woolf, , and hethttp://www.ctu.dk/tsa/) was used for TSA analysis.In an attempt to include all the eligible trials in this meta-analysis, and to minimize the systematic errors (bias) or random errors by chance, we used a novel statistical TSA tool from Copenhagen Trial Unit, Center for Clinical Intervention Research, Denmark. TSA estimates the required information size and adjusts statistical significance thresholds, and also estimates the power of conclusion  statistics for each of the (29) cases considered in our study. CA trend test is the most popular association analysis for determining genetic associations Sasieni, .CA trend test statistic (T) is given by the following equationWhereM1 = 0; M2 = 1;M3 = 2.For a codominant or additive model, M1 = 0; M2 = 1; M3 = 1.For a Dominant Model, M1 = 0; M2 = 1; M3 = 2.For a Recessive model, Ny is the total number of controls, Nx is the total number of cases, Xi and Yi represents the number of cases and controls for each of the three  genotypes.Here, During the study search the full-texts of all the articles deemed potentially eligible were retrieved. The study eligibility for inclusion was determined by reviewing the full text of all the articles by the first investigator. The second investigator randomly selected 10% of the articles and reviewed them by the same procedure, independently. A complete agreement was observed between the two investigators regarding inclusion and exclusion of the studies. Following the identification of the final set of the eligible articles, one investigator extracted the relevant data from all the studies, and the other investigator independently re-extracted the data from all the included studies to cross-check this step. Table p < 0.05 was fixed for the significant publication bias in the present meta-analysis. All the comparison genetic models showed absence of publication bias (p > 0.05)  Table .Q-test and I2 statistics among the included studies. Four genetic models showed heterogeneity, so the data was synthesized by applying the random effect model , homozygous , dominant  and recessive   genetic model did not show any SLE risk  genetic model. Similarly, dominant  genetic model also showed marginal significant risk. But, other genetic models, i.e., homozygous , heterozygous  and recessive  genetic models did not show any increased or decreased risk of SLE , homozygous , heterozygous , dominant  and recessive  genetic models , homozygous , heterozygous , dominant  and recessive  genetic models  was reached. The dominant model was taken as an example in the TSA analysis, which indicated that d Figure . Subgroud Figure .p-values can be referred from the attached Excel Sheet: A comparison of the results obtained from OR and CA trend test is shown in Table Earlier studies have reported that individual's SLE susceptibility often present a variety of symptoms that pose difficulty in the disease diagnosis by the physicians, leading to a consequent delay in the onset of the treatment. Host factors, including genetic polymorphisms implicated in autoimmune diseases, might have interpreted this divergence. Thus, there is an urgent need of the identification of the genetic biomarkers that are responsible for the onset and progress of SLE. Nowadays, the current research focus is based on the role of genetic susceptibility in autoimmune disease progression or development.ACE plays an important role in SLE etiology by affecting the immune responses and vascular changes. Angiotensin II converted by ACE is a potent pro-inflammatory modulator for immune responses in the renal tissues and shows robust potential in mediating the development and progression of renal disease during SLE  genotype/allele might confer susceptibility in SLE patients. Individual studies generally have low statistical power to detect the risk, owing to their small sample sizes. Therefore, it is more judicious to estimate the precise relationship of ACE I/D gene polymorphism to understand the contribution of this polymorphism in overall SLE risk.Renin—angiotensin system (RAS) regulates the arterial blood pressure at both levels—systemic and tissue, and contributes to the immunological responses arising during various phases of nephropathy evolution Egido, . The majACE I/D polymorphism with an increased SLE risk. These results confirm that ACE I/D genetic variant may interfere with its expression level and plays a pivotal role in the progression of SLE. Earlier studies have reported that circulating ACE levels vary greatly between individuals and is extremely determined genetically. ACE I/D polymorphism could alter the circulation of ACE levels. It has been found that individuals carrying DD genotype had 2-folds higher circulating ACE levels and lowest among the II genotype carrying individuals  was further confirmed by the TSA and comparison with CA trend test statistics, which authenticated the association of Above studies have reported higher serum ACE levels in SLE patients. However, the interruption of the renin angiotensin with ACE inhibitors or angiotensin receptor blockers (ARBs) is recommended as first line adjuvant therapy for the patients with lupus nephritis for proteinuria  for increased statistical power and robust conclusion; well exploration of the methodological issues generally occurring in pooled analysis  which further confirmed the reliability and validity of the present study; and reduction of type I error rate by Trial sequential analysis.In comparison with the previously published meta-analyses , this study also suffers with several limitations, which must be stated and addressed in the future studies. First, the result of overall population may be biased, perhaps owing to slight over-representation of Asian population. Second, the result of subgroup analysis especially for Caucasian and African, it may be possible that the observed result could be partly due to modest number of studies included. Third, there was a significant heterogeneity in some of the pooled analysis, which may affect the meta-analysis outcome. Fourth, unadjusted estimates form the basis of this meta-analysis and studies published in English language only were included. Unpublished data and ongoing studies were not searched and included. Further, studies reporting negative findings are less likely to be published, hence causing publication bias thereby bringing increase of the associations. Fifth, due to lack of sufficient data, we were not able to study lupus nephritis individually in relationship with ACE I/D gene polymorphism and SLE susceptibility. Individuals carrying allele “D” of the ACE I/D polymorphism had greater risk of SLE. The findings will advance our understanding of the role of ACE I/D genetic variant and assist in identifying the “at-risk” individuals. Our results also provide a solid foundation for future genetics studies to focus on ACE related phenotypes and integrative network modules analysis to clarify the potential role of ACE genetic variants in SLE risk. Moreover, more meta-analysis with larger sample size should also be encouraged in future to understand the molecular mechanism of ACE gene with SLE development and to verify the current findings reported in this manuscript.In conclusion, the pooled results of independent association studies by meta-analysis confirmed statistically significant association between SK, SD, RM, MW, AJ, AP, ML BM, NA, and SH conceived and designed the study and experiments. SD, SK, RM, MW, AP, NA, and SH performed the experiments. SD, SK, RM, AJ, AP, ML, BM, and NA analyzed the data. AJ, MW, AP, BM, ML, NA, SD, and SH contributed reagents, materials, analysis tools. SD, SK, RM, AP, and SH wrote the paper. All authors reviewed and approved the manuscript.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.


Objective: A number of studies have reported that aldehyde dehydrogenase-2 (ALDH2) polymorphisms maybe associated with the risk of Alzheimer's disease (AD) and Parkinson's disease (PD). However, the results of such studies are inconsistent. We therefore conducted a meta-analysis to clarify the association between ALDH2 polymorphisms and the risk of AD and PD.Methods: Five online databases were searched and the relevant studies were reviewed from inception through May 10, 2018. Odds ratios (ORs) and corresponding 95% confidence intervals (CIs) were calculated in each genetic model of the general population and various subgroups. Furthermore, we simultaneously performed heterogeneity, cumulative, sensitivity, and publication bias analyses.Results: Overall, nine case-control studies involving 5,315 subjects were included in this meta-analysis. Potential associations were found between the ALDH2 rs671 G>A polymorphism and the risk of AD . In addition, some similar results were observed in other subgroups. Moreover, no significant association between ALDH2 polymorphisms and PD risk.Conclusions: In conclusion, our meta-analysis indicated that the ALDH2 rs671 G>A polymorphism plays an important role in AD development. Neurodegenerative disorders are a family of heterogeneous disorders, in which progressive degeneration occurs in the structure and function of the central or peripheral nervous system . NeurodeAD and PD are closely related to aging, can have multiple causative factors, such as oxidative damage, abnormal protein deposition, and neuroinflammation . Other fALDH2) gene. This gene is located on chromosome 12q24.12, comprises 13 exons and 12 introns, and encodes an important biologically active enzyme, ALDH2. This enzyme participates in the metabolism and detoxification of aldehyde, and it can metabolize short-chain aliphatic aldehydes and converted acetaldehyde into acetate. Moreover, ALDH2 is involved in the metabolism of other biogenic aldehydes, such as 4-hydroxynonenal, 3,4-dihydroxyphenylacetaldehyde, and 3,4-dihydroxyphenylglycoaldehyde  of rs671, inherent in mainly the East Asian population, deregulates the ALDH2 activity intrinsically, and induces the accumulation of acetaldehyde. In 2000, Kamino et al. conducted the first case-control study in the Japanese population and found that the A allele and GA/AA genotypes are associated with an increased risk of AD :Five online databases  were used to search for related studies on the association between Aldehyde dehydrogenase 2#1 Aldehyde Dehydrogenase-2#2 ALDH2#3 #4 rs671#5 rs4767944#6 rs441#7 #1 OR #2 OR #3 OR #4 OR #5 OR #6#8 polymorphism#9 variant#10 mutation#11 #8 OR #9 OR #10#12 neurodegenerative disorders#13 Alzheimer's disease#14 Parkinson's disease#15#12 OR #13 OR #14 OR #15#16 #7 AND #11 AND #15.ALDH2 polymorphisms and the risk of AD and PD; (2) studies containing sufficient data on the genotype in the control groups to evaluate crude odds ratios (ORs) and 95% confidence intervals (CIs); (3) studies published in English or Chinese, and (4) if overlapping or duplicate data were found on the same theme, only the largest or most recent sample data were included. The exclusion criteria included: (1) case report or review articles; (2) molecular biology research; (3) studies without efficient data; and (4) studies with duplicated or overlapping data.The following were our inclusion criteria: (1) observation studies focusing on the association between Two authors (Chen and Huang) independently reviewed the included studies, and the following information was extracted and recorded for analysis: the first authors' name, publication date, study country, control design, genotyping method, sample sizes of the cases and controls, frequency data for the genotype distribution, assessment of Hardy-Weinberg equilibrium (HWE) in control, minor allele frequency, and disease type. The modified Newcastle-Ottawa scale (NOS) was used to evaluate the quality of all included studies . The scoALDH2 polymorphisms and the risk of AD and PD. For example, the following five genetic models of the rs671 G>A locus were used: allele contrast (A vs. G), co-dominant models (GA vs. GG and AA vs. GG), dominant model (GA+AA vs. GG), and recessive model (AA vs. GG+GA). Heterogeneity among the included studies was examined using Cochran's Q tests and I2-tests . Subgroup analyses also revealed an increased AD risks in Chinese  and Japanese  on the basis of country difference. The subsequent analysis based on the HWE status and other subgroup revealed the similar increased associations .Association between the rs671 polymorphism and the risk of AD and PD. There were six case-control studies involving 2,840 subjects focused on the association between the rs671 polymorphism and AD risk. The aggregated results indicated an increased risk of the rs671 G>A polymorphism in patients with AD .Publication biases were investigated, and the results did not show any obvious asymmetry in the five funnel plots , and this trend was observed in both males and females. Wang et al. found an apparently increased AD risk in Chinese individuals with the rs671 A-allele . Similarly, Zhao et al. and Ma et al. reported the similar elevated risk for PD (Since 2000, an increasing number of studies have focused on the relationship between k for PD  and AD wk for PD . HoweverALDH2 rs671 G>A polymorphism and the risk of AD and PD. All results suggest that the polymorphism locus of ALDH2 rs671 G>A may be a potential risk factor for AD but not for PD in the East Asians. The current evidence indicated that the carriers with AA genotype were more dangerous compared with the GG genotype, which was also coincides with the decrease of ALDH2 protein activity that caused by allele A mutation. Moreover, the elevated risks we identified were also observed in some subgroups, such as in male groups. Among East Asians, males account for the majority of people who consume alcohol. The interactive effects of alcohol consumption and the deficient ALDH2 enzymatic activity caused by the rs671 mutation may contribute to AD development in males. Two of the studies we evaluated focused on the rs4767944 polymorphism, and the results revealed a slightly protective effect of this polymorphism against PD. In addition, no positive association was identified between rs441 polymorphism and the risk of PD with only one study. However, given the limited number of studies and included participants for this polymorphism, the results might not reflect the real relationship. Therefore, additional studies examining the association between the rs4767944 and rs441 polymorphisms and the risk of PD are necessary.To our knowledge, this is the first meta-analysis on the association between the ALDH2 rs671 G>A polymorphism, but not for the rs4767944 and rs441 polymorphisms. The investigations of the associations of each of the three polymorphisms with the risk of AD and PD were conducted independently without adjusting for gene-gene interactions, such as through a haplotype analysis. Third, the interactions with some factors, such as body mass index, blood pressure, and unhealthy living habits, were not examined because of the lack of accurate individual information in the included studies. Finally, heterogeneity was observed in some of the genetic models in all included studies, and the meta-regression failed to identify any factors that contributed to the heterogeneity. Future studies that address these limitations will be required before any concrete conclusions about the relationships between these polymorphisms and AD and PD can be made.Given the inherent deficiencies in meta-analyses, this study has some limitations that should be considered when interpreting our findings. First, all subjects in the included studies were from East Asian countries. As such, the pooled results of this meta-analysis only reflect the East Asian population and thus are not generalizable to other ethnicities. Second, a slightly large sample size was only available for the ALDH2 rs671 G>A polymorphism may be a potential risk factor for AD. Additional case-control studies are needed to investigate the underlying mechanism of the potential risk.In conclusion, the present results indicate that the JC, WH, and LZ conceived the study and wrote the draft of the paper. JC, WH, and Y-YH searched the databases and extracted the data. C-HC and G-BJ analyzed the data and reviewed the manuscript. All the authors approved the final manuscript.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.


P > 0.05), there was no evidence of significant publication bias. Moreover, our DR analysis indicated that the risk of NHL grew by 26% for each additional microgram of nitrite consumed in the diet per day . Through subset analysis of the nitrite studies, data from the high-quality studies indicated that consumption was positively associated with carcinogenicity, leading to NHL  and positively correlated with the development of diffuse large B-cell lymphoma , but not other NHL subtypes. In addition, the data suggested that females  and high levels of nitrite intake  had a higher risk of NHL. Our meta-analysis supports the hypothesis that nitrite intake, but not that of nitrate, raises the risk of developing NHL. In the future, better designed prospective research studies should be conducted to confirm our findings, clarify potential biological mechanisms and instruct clinicians about NHL prophylaxis.Epidemiologic studies of the relationship between nitrite or nitrate consumption and risk of non-Hodgkin lymphoma (NHL) remain controversial. The current meta-analysis aimed to reexamine the evidence and quantitatively evaluate that relationship. Manuscripts were retrieved from the Web of Science, Chinese National Knowledge Infrastructure and PubMed databases up to May 2019. From the studies included in the review, results were combined and presented as odds ratios (OR). To conduct a dose-response (DR) analysis, studies presenting risk estimates over a series of categories of exposure were selected. Our data indicate that the consumption of nitrite was linked to a significantly increased hazard of NHL , rather than nitrate . According to Egger’s and Begg’s tests ( According to the 2018 global cancer statistics, NHL ranks as the 8th most common carcinoma in males and 10th in females2. In the USA, it has been estimated that 74,680 new cases of NHL  and 19,910 deaths  occurred in 20183. Despite the diagnostic and therapeutic progress in the recent decades7, the 5-year survival rate for all NHL forms combined is 72% (https://seer.cancer.gov/statfacts/html/nhl.html). It is well-known that NHL patients, particularly for high-risk subtypes , continue to receive adverse prognoses. Hence, to increase precautions against and reduce the prevalence of NHL, further exploration of its risk and deeper understanding of its epidemiology are essential.Non-Hodgkin lymphoma (NHL) is a heterogeneous group of hematologic malignancies, developing from cells contained in the lymphoid tissue or from lymph glands. In recent decades, NHL has been categorized into more than 40 forms based on pathological and histological features by the World Health Organization8. In epidemiological studies, the potential relationship between nitrate or nitrite consumption and risk of development of tumors has been investigated. Positive associations have been detected in adult glioma 9, thyroid neoplasm 10, and gastric carcinoma 11, etc. Moreover, Xie et al.12 and Song et al.13 further confirmed these correlations by conducting meta-analyses. During the past decades, the relationship between risk of NHL and nitrate or nitrite consumption have been investigated in several epidemiological publications, yet the results are contradictory. Some surveys failed to show positive or negative associations21, while others have revealed significant correlations25. When no specific trend can be ascertained from any individual investigation, combining a number of independent studies can reveal hidden associations through meta-analysis. Therefore, we conducted the present meta-analysis to discover latent connections between nitrate or nitrite consumption with the etiology of NHL, and so aimed to: (1) evaluate the epidemiological evidence about the relationship between nitrate or nitrite consumption from water or diet and risk of NHL; (2) consider a possible dose response (DR) relationship between nitrate or nitrite consumption and the risk of NHL; and (3) assess the quality of evidence and the statistical significance of the results.Nitrate and nitrite are crucial precursors of N-nitroso compounds (NOCs), a group of genotoxic composites that are highly carcinogenic and which act systemically25.Figure 25. In each study included in the review, at least two different pathologists were involved in the diagnosis of NHL. Every NHL case was categorized by the Working Formulation criteria of the Lymphoma Study Group or using the World Health Organization (WHO) classification criteria27. The quality of each study was evaluated using the Newcastle-Ottawa Quality Assessment Scale (NOS). Scores ranged from 4 to 8 with a mean value of 6   , with both Egger’s and Begg’s tests indicating that no evidence of publication bias existed   and risk of NHL is illustrated in Fig. 83) Fig. . No evidly) Fig. . Furtherly) Fig. . Conversly) Fig. . No signly) Fig. . Similarly) Fig. .Figure 2Subset analyses were subsequently conducted, based on study design or quality, gender, source of nitrate, geographical region, NHL subtypes or levels of nitrite or nitrate Tables  and 3. IBecause a heightened risk of NHL caused by the intake of nitrites, but not nitrates, was found, a DR analysis was further conducted to assess the dose-response interrelationship for the risk of NHL and nitrite consumption. A 26% greater risk of NHL was linked to an increase of 1 microgram of nitrite per day  , but not studies of nitrate  . Furthermore, the overall association was more robust . As displayed in Table et al., as can be seen from the Galbraith plot . The pooled OR remained not significant . As shown in Table In order to assess heterogeneity in the studies of this meta-analysis, the lot Fig. . After eet al.24 and the research by Ward et al.25, respectively. For nitrate intake and risk of NHL, the ORs of the eleven studies varied from a high of 1.02 (0.95–1.11) to a low of 1.00 (0.91–1.09) after omitting the studies of Ward et al.25 and Chang et al.20, respectively.The influence of each study on overall estimate of risk was measured by repeating the meta-analysis after omitting each study in turn so as to conduct an analysis of sensitivity. As shown in Supplementary Fig. etc.), nitrites and nitrates may be adjuncts that maintain the color of meat products and reduce microbial spoilage. However, excessive intake of processed meat is associated with an elevated risk of carcinomas, including NHL28. In recent years, the association between risk of NHL and nitrate or nitrite intake have been explored in a number of epidemiological studies. However, the relationship remains uncertain. Therefore, all available studies that have explored these relationships were integrated into this meta-analysis in an attempt to clarify these possible associations.Globally, nitrates and nitrites are present within many typical diets. For example, in processed meats . In summary, the results suggest that nitrite intake, but not nitrate, is an important factor in the occurrence of NHL and that increased nitrite intake appears to raise the risk of NHL in a dose-dependent manner.29. In fact, this meta-analysis was in agreement with previous studies, the results of which indicate that nitrate consumption was not linked to increased risk of NHL. Moreover, in subtype analysis where study quality or design, geographical area, source of nitrate, NHL subtypes, levels of nitrate in water and gender were considered, the risk of NHL was not increased as a result of nitrate consumption. Indeed, in representative exposure models, several studies have provided strong evidence that nitrate cannot be a carcinogen for humans or animals31. Notably, Palli et al., Rogers et al. and Ward et al. separately demonstrated that high nitrate consumption protected against esophageal cancer, gastric cancer and renal cell carcinoma34. However, high nitrate intake was deemed to be from the consumption of vegetables, which are also rich in various anti-cancer substances  declared, as early as 2010, that there was no conclusive evidence that nitrates were carcinogenset al. observed that nitrite intake promoted lymphomas in rats in a dose-dependent manner35. In agreement with this study, our results indicated a significant positive relationship was found between NHL and high levels of nitrite intake , but not for low levels of nitrite intake . However, the mechanisms by which nitrites could influence the pathogenesis of NHL require clarification. A number of potential mechanisms could be responsible for nitrites being a risk factor for NHL. Previous studies have reported that nitrites can be converted into NOCs in the stomachs and intestines of both humans and animals through nitrosation with amides and amines. It is generally believed that NOCs are among the strongest known carcinogens, including N-nitrosodiethylamine (NDEA), N-nitrosopiperidine (NPIP), N-nitrosopyrollidine (NPYR), N-nitrosodimethylamine (NDMA) and N-Nitrosodibutylamine (NDBA), etc36. They are considered genotoxic procarcinogens associated with driving tumorigenesis in a variety of tissues in more than 40 species, even senior primates38. In agreement with previously published reports, Storer et al.39 found that NDEA was able to effectively increase the frequency of malignant lymphoma in Eμ-pim-1 transgenic mice. Additionally, through the interplay of superoxides and NO production nitrites are able to form peroxynitrites which have powerful oxidative capability that causes damage to DNA through cellular oxidation, closely associated with carcinogenesis41. In vitro, Ustyugova et al.42 demonstrated that nitrites could inhibit Th1 cytokine formation, including interferon-γ, tumor necrosis factor-β and interleukin-2. Interestingly, Saberi et al. confirmed that such a change in concentration of these cytokines, at least partly, contributes to the increased risk of developing NHL43. In our study, a significant positive relationship was showed between the risk of NHL and nitrite consumption for female , but not for male . The different hormone and cytokine levels in female and male might be a possible reason for raising the risk of NHL. A greater number of basic studies are urgently required in order to clarify the biological mechanisms in carcinogenesis induced by nitrites in NHL.In 1979, Newberne The present meta-analysis is the largest study to date in which the relationship between nitrate and nitrite consumption from diet or drinking water has been studied in relation to the risk of NHL, including 8,067 NHL cases in total. Based on the studies included in this review, it was possible for us to investigate the relationship in various subgroups using meta-analysis methods. Nevertheless, the meta-analysis had some limitations, due to data originating from previously published observational studies. Firstly, our analysis assembled together published studies in Chinese or English but did not attempt to uncover unpublished data, which may have led to publication bias. However, the Begg’s and Egger’s tests did not suggest any apparent evidence for recall bias. Secondly, in this meta-analysis, the results were principally derived from case-control studies, which could lead to the potential for publication bias, owing to the data for case-control studies being retrospective. Thirdly, due to a number of the subgroup analyses being conducted on tiny datasets, a high degree of confidence should not be placed on their conclusions. Fourthly, although we found a positive relationship between nitrite consumption and DLBCL, we did not obtain sufficient data to compute odds ratios for other particular histopathological subclasses of NHL from the studies included in this meta-analysis. Therefore, we were not able to explore the relationship between nitrate or nitrite consumption and the risk of suffering additional subclasses of NHL. Fifthly, when exploring the relationship between nitrate intake and the risk of NHL, we combined the data for nitrate consumption from diet and drinking water. However, the metrics for nitrate in diet or drinking water was different and we could not find an accurate method for combining them, which would affect the reliability of this meta-analysis. Additionally, a wide range of values for the cutoff points for the highest and lowest level of the consumption of nitrate and nitrite was found in the involved studies, which could led to the possible bias. Finally, the accuracy of the conclusions of this meta-analysis will have been influenced by the threshold values of the highest and lowest categorizations of nitrate or nitrite consumption being distinct in the various studies. Hence, a greater number of well-designed, multi-center, large-sample epidemiological studies are essential for better elucidating the association between the risk of NHL and nitrate or nitrite consumption.To summarize, the results suggest that nitrite intake is linked to increased risk of NHL. In the future, to acknowledge our conclusions and to ensure precautions against NHL, additional and more stringent systematic studies are required.45.A systematic and comprehensive article retrieval strategy that provided a general impression of the risk of NHL due to nitrate or nitrite consumption was conducted. The Web of Science, the Chinese National Knowledge Infrastructure (CNKI) and PubMed databases were searched for articles of follow-up, cohort or case-control studies assessing the relationship between the risk of NHL and nitrate or nitrite consumption from drinking water or diet, from inception of each database until May 31th, 2019. Few relevant articles were obtained by searching for the terms ‘nitrate’ or ‘nitrite’ and ‘non-Hodgkin lymphoma’. Identification of more relevant articles was accomplished by combing the keywords in a more detailed retrieval strategy, as follow: (N-nitroso compounds OR nitrite OR nitrate) AND (NHL OR non-Hodgkin’s lymphoma OR non-Hodgkin lymphoma OR lymphoma). In addition, a manual search of the references of relevant articles was performed to locate additional studies not identified in the initial search. The current analysis was performed with due consideration to the quality criteria for meta-analysesAll relevant articles were included if they fulfilled the following criteria: (1) a follow-up study or the study had a case-control design; (2) the study investigated the possible relationship between risk of NHL and nitrate or nitrite intake from diet or drinking water; (3) outcomes included relative risk (RR) or odds ratio (OR), or provided sufficient data to perform calculations; (4) the study was published before May 2019 and reported in Chinese or English. If the same samples were reported in different articles, only the manuscript reporting the largest sample size was selected. If an article provided inadequate details or reported overlapping material, it was precluded.The following data were retrieved from each article: article title (including publication year and first author’s name), study location of origin, judgment of risk factors, numbers of cases and controls, study design, adjustment factors and patient country of origin. Because non-Hodgkin lymphoma is an uncommon disorder, it is believed that the RR is commensurate with the OR. Thus, OR was adopted to assess any potential association between risk of NHL and nitrate or nitrite consumption. And OR was calculated by comparing the highest level of nitrate or nitrite consumption with the lowest. Two researchers (C.H. and M.X.Y.) independently conducted data collection via a structured questionnaire. A third researcher was consulted in order to reach a consensus, if required.46. In the event of disagreement between the two investigators, a third reviewer arbitrated. The NOS is a tool that permits appraisal of follow-up, population selection, exposure and comparability of included studies. It was then possible to quantitatively assess the quality of each article according to the four aspects above. For each study, the points total varied from 9 to 0. Scores ≥7 represented articles considered high-quality, while scores <7 were considered of low quality.Two investigators independently evaluated the quality of the articles included in the review through application of the nine-star Newcastle-Ottawa Scale (NOS)47. Conversely, the DerSimonian and Laird method was used to evaluate a random-effects model48. In order to calculate levels of heterogeneity, the I2 statistic  and the Q test were adopted49. To detect sources of heterogeneity, we also conducted subgroup analyses, using source of nitrate, study design or quality, geographical area, gender, NHL subtype and levels of nitrate in water. In addition, a Galbraith plot and meta-regression analysis were also performed to explore the potential origins of heterogeneity50.The relationship between risk of NHL and nitrate or nitrite intake was evaluated using pooled ORs and their 95% confidence intervals (CIs). The choice of random or fixed-effects model was dependent on heterogeneity between studies. Where significant heterogeneity was absent, a Mantel-Haenszel test was utilized to compute pooled ORs in a fixed-effects modelet al.51 and Greenland et al.52, we repeated the meta-analysis to re-assess risk. The mean value of every categorization range for nitrate or nitrite intake was used in the DR analysis. If the upper boundary of the highest category or the lower boundary of the lowest category was not provided in a particular study, the equivalent interval was assumed according to the next highest or lowest category53. The potential DR relationship of risk of NHL to nitrate or nitrite intake was estimated in two stages51. In the first, a limited cubic spline model with 4 knots at the percentages: 5%, 35%, 65% and 95% of the allocated exposure consumption was calculated, then three regression moduli (4 knots minus 1) were computed. Secondly, the covariance from every study was integrated.For DR analysis, articles were included if they provided the number of controls and cases in addition to reporting at least 3 levels of nitrate or nitrite intake for every exposure classification. Using a method proposed by Orsini 54, which indicated the magnitude of influence of all studies allocated normally around the center of a funnel plot. In addition, latent publication bias was also evaluated using an Egger’s test (linear regression method)55 and a Begg’s test (rank correlation method)56. STATA 11.0 software  was used to conduct the meta-analysis. P values less than 0.05 were considered significant.In order to assess underlying publication bias, we executed a trim-and-fill testsupplementary information.


MDM2 40bp indel polymorphism and cancer susceptibility. Eligible studies were retrieved by searching Web of Science, PubMed, Scopus, and Google scholar databases up to August 27, 2018. The pooled odds ratios (ORs) with 95% confidence intervals (CIs) were calculated to estimate the strength of association between the polymorphism and cancer risk. The findings of this meta-analysis revealed that the 40bp indel polymorphism significantly increased the risk of overall cancer risk in heterozygous  and ID+DD  genotypes. Stratified analysis by cancer type proposed that the study indel variant significantly associated with the risk of gastrointestinal cancer in heterozygous  and ID+DD  genotypes. The present findings showed a significant association between the MDM2 40bp indel polymorphism and overall cancer risk as well as gastrointestinal cancer susceptibility. Larger and well-designed researches are required to validate the findings association in detail.This meta-analysis aimed to provide an up-to-date comprehensive evaluation on the association between the  MDM2, OMIM: 164785) gene is mapped to 12q14.3-15 [MDM2 have been shown in various cancer types [) in the MDM2 promoter P1 region, may alter the expression of MDM2 [MDM2 40bp indel polymorphism and the risk of various cancers [Cancer remains one of the main leading cause of morbidity and mortality and poses a serious challenge to global public health worldwide . Cumulater types , 10. Gener types -13. A 40 cancers , but theLiterature search: We performed a comprehensive search for relevant studies focusing on MDM2 40bp indel polymorphism in PubMed, Web of Science, and Scopus databases up to November 02, 2018. The search keywords were “cancer or tumor or carcinoma or neoplasms” and “MDM2 or mouse double minute 2” and “polymorphism or mutation or variant or deletion or indel or rs3730485 or del1518”. Relevant studies comprised the meta-analysis if they met the following inclusion criteria: 1) Original case-control studies; 2) studies provided sufficient genotyping data of MDM2 40bp indel polymorphism in both cases and controls. The exclusion criteria were: 1) case reports, conference abstract, meta-analysis, and duplication data; 2) studies lacking genotype information.Data extraction: Two investigators independently searched the databases and extracted the relevant data from eligible studies. The following data was recorded from each study including the first author, Year of publication, country, ethnicity, source of control, cancer type, genotype distributions in cases and controls and result of the Hardy-Weinberg equilibrium (HWE) test (WE) test .Statistical analysis: All analyses were done by STATA 14.1 software . Departure from HWE in controls was examined by the chi-square test. The strength of the association between MDM2 40bp indel polymorphism and cancer risk was evaluated by pooled odds ratios (ORs) and their 95% confidence intervals (CIs). The Z-test was used for statistical significance of the pooled OR. We estimated the between-study heterogeneity by the Q-test and I2 test. The p<0.10 indicating the presence of heterogeneity. If heterogeneity exist, a random-effect model was employed; otherwise, a fixed-effect model was used. Stratified analyses by cancer type was also applied for each genetic comparison model. We assessed publication bias visually using funnel plots and conducting quantitative estimations with Egger’s and Begg's tests. Sensitivity analysis was executed by removing each study time to inspect the impact of individual data set on the pooled ORs.A flow chart of the study selection process is shown in The main findings of our meta-analysis and the heterogeneity test are presented in Between-study heterogeneity across studies included in the analysis is shown in Sensitivity analysis was done to evaluate the stability of the findings in our meta-analysis. The sensitivity analysis revealed no obvious effects from each study in homozygous codominant, and recessive genetic models.MDM2 polymorphism may modulate the risk of cancer. In the present study we conducted an updated meta-analysis to find out the correlation between the 40bp indel polymorphism of MDM2 and cancer risk. Fourteen independent article [MDM2 40bp indel polymorphism on cancer risk were pooled in this analysis. In our meta-analysis, 5 genetic models were considered including homozygote codominant, heterozygous codominant, dominant, recessive, and allele to evaluate the impact of MDM2 40bp indel polymorphism on cancer risk. The overall analysis revealed that heterozygous codominant, and dominant increased the risk of cancer. Subgroup analysis by cancer types proposed that MDM2 40bp indel polymorphism increased the risk of gastrointestinal cancer in heterozygous codominant, and dominant genetic models. No significant association was observed between the variant and the risk of breast cancer, ESCC, lung cancer, prostate cancer, and ovarian cancer, which may be due to the small number of articles.The tumor suppressor p53, a transcriptional factor, really controls the growth and development of normal cells. P53, serves as an important tumor suppressor protein in preventing cancer, regulates the cell cycle and apoptosis -31. Give article -28 incluMDM2 40bp indel polymorphism on cancer susceptibility. They found lack of association between this polymorphism and cancer risk. One of the study they enrolled in the meta-analysis was not related to cancer [Recently, Hua et al  publisheo cancer . In addio cancer . The degree of heterogeneity is an essential factor assessed in genetic association meta-analysis. In our meta-analysis, the genetic models which associated with cancer risk showed no evidence of heterogeneity. Furthermore, assessment of publication bias showed no obvious publication bias in the funnel plot under all genetic models in overall cancer as well as gastrointestinal cancer. After omitting each study in order, the pooled ORs of the remaining studies were comparable to the total pooled ORs in homozygous codominant and recessive genetic models, suggesting that the meta-analysis was stable.Several limitations of our meta-analysis should be taken into account. First, only studies published in English were selected. Second, heterogeneity existed among the included studies. Although, the sources of heterogeneity were not clear, it may be derived from differences in cancer types and ethnicities. Third, the sample size of our meta-analysis was still relatively small in stratified analysis by cancer types . So, the statistical power was limited. MDM2 40bp indel polymorphism is a risk factor for developing overall cancer as well as gastrointestinal cancer. More well-designed large-scale case-control studies are necessary to elucidate the possible roles of this variant in cancer.Despite the limitations, our meta-analysis suggest that


Using a conventional method to assess the significance of meta-analysis , statins had a statistically significant effect on reducing cancer incidence in 10 of 18 types of cancer. When we graded the level of evidence, no cancer type showed convincing evidence, and four cancers  showed suggestive evidence of a preventive effect. There was weak evidence of an association with six cancers, and no significance for the remaining eight cancers. None of the meta-analyses of RCTs on the association of statin and cancer incidence showed a statistical significance. Although there was a preventive effect of statin on cancer incidence in 10 of the 18 cancer types, the evidence supporting the use of statins to reduce cancer incidence was low. Therefore, the associations between statin use and cancer incidence should be carefully considered by clinicians.Statins are reported to reduce the risk of cancer, but the results of various published studies have been contradictory. We carried out an umbrella review to provide an overview and understand the strength of evidence, extent of potential biases, and validity of claimed associations between the use of statins and cancer incidence. We comprehensively re-analyzed the data of meta-analyses of randomized controlled trials (RCTs) and observational studies on associations between statin use and cancer incidence. We also assessed the strength of evidence of the re-analyzed outcomes, which were determined from the criteria including statistical significance of the  Cancer places one of the biggest burdens on health care system in both highly developed and less developed countries, and its incidence and mortality have been increasing for decades mainly due to longer life expectancy . Based oStatins, competitively inhibiting 3-hydroxy-3-methylglutaryl coenzyme A reductase, have been used for lowering cholesterol levels ,5. BesidTo understand and evaluate the strength of the evidence of the effect of statins on reducing cancer incidence, we carried out an umbrella review and comprehensively re-analyzed the data of meta-analyses of randomized controlled trials (RCTs) and observational studies.We performed an umbrella review of meta-analyses and systematic reviews reporting on the associations between statin use and the incidence of cancer. This umbrella review was conducted according to the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) guidelines . The PRIWe searched the PubMed database and limited the articles to those written in English, regardless of the publication date. The final search was performed in August 2018. The keywords we used were the following: ‘(hydroxymethyl glutaryl-coa reductase inhibitor OR statin) AND (cancer OR neoplasm OR tumor) AND ’. Meta analyses of either RCTs or observational studies were included in our search strategy. We reviewed the retrieved articles by examining the titles, the abstracts, and the full texts, then decided which article to include or exclude. We further searched the EMBASE database for potentially eligible meta-analyses, but no additional meta-analysis was included because the identified meta-analyses were lacking data necessary for performing re-analysis or overlapped with the PubMed search. The detailed search strategy is presented in We included meta-analyses and systematic reviews of both RCTs and observational studies reporting on the relationship between statin use and cancer incidence. Observational studies included both cohort and case-control studies. We excluded review articles without meta-analysis, in vitro studies, and genetic studies. We also excluded meta-analyses lacking data necessary for performing re-analysis. If an article presented more than one meta-analysis, all meta-analyses were included and assessed separately by study design or cancer type.Two investigators (G.H.J. and J.I.S) independently extracted the data, and discrepancies were resolved through consensus. We obtained the data from eligible meta-analyses and extracted and summarized the information on first author, year of publication, the type of cancer, the study design, the number of included studies, the number of cancer cases and total participants, and the random effects with a 95% confidence interval (CI). From the eligible studies, we also extracted the raw data of each individual study for further meta-analysis by combining all data by cancer and study design. If a single study consisted of both RCTs and observational studies, we separated the studies according to the study types  and reported the results separately.p-value with both random- and fixed-effects. All re-analyses in this study were performed using the Comprehensive Meta-Analysis software ver.3.3.070 .We firstly re-analyzed each meta-analysis and reported the relationship between statin use and cancer incidence. In addition, if there were overlapping meta-analyses on the same topic, we combined all the individual studies from eligible meta-analyses according to the type of cancer and study design and performed a re-meta-analysis after eliminating overlapping individual studies and including missing individual studies. We presented the summary effect size, 95% CI, and For each meta-analysis, we re-analyzed the individual studies and estimated the summary effects and 95% CI using both random- and fixed-effects methods . We alsoI2 metric of inconsistency and the p value of the X2-based Cochrane Q test. I2 values of <50%, 50%–75%, and >75% are usually judged to represent low or moderate, large and very large heterogeneity, respectively [We assessed heterogeneity across the studies using the ectively .p-value < 0.10) in random-effects meta-analyses, we decided that the study has small-study effects.Publication bias was evaluated by using Egger’s regression test . Small sp-values, 95% PI, a small-study effects, a between-study heterogeneity, and concordance between the effect estimate of the largest study and summary estimate of the meta-analysis [We determined the level of evidence of each meta-analysis and re-analyzed the pooled meta-analysis to classify the strength of the evidence of the association between statin use and cancer incidence. The criteria were set according to the statistical significance by random and fixed-effects analysis . The crip-values at p < 0.001. No small study effects or large between-study heterogeneity were found, and 95% PI rejected the null hypothesis. There was a concordance between the effect estimate of the largest study and the summary effect of the random-effects meta-analysis.Convincing evidence: There was a statistical significance for the random-effect and fixed-effect p < 0.05, but a 95% PI included the null hypothesis. No small study effects or large between-study heterogeneity were found.Suggestive evidence: There was statistical significance of random effects at p < 0.05. Small study effects or large between-study heterogeneity were found.Weak evidence: There was a statistical significance of random effects at p > 0.05)Non-significant association: There was no statistical significance by random effect meta-analysis , while eight associations  were not significant  were statistically significant under the conventional interpretation of meta-analysis criteria  and carried out a re-meta-analysis of the pooled raw data in association with statin use and cancer incidence . All oveTable 3Of the 18 types of cancer, three cancer types  did not have meta-analyses using RCTs. Among the other 15 cancer types, there was no statistically significant statin–cancer incidence association in meta-analyses of RCTs . For thep-value, 10 of 18 associations of cancer incidence showed a statistically significant preventive effect of statin.The purpose of this umbrella review of previous meta-analyses and re-analysis of meta-analyses, including all the individual studies, was to highlight the potential effects of statin use on cancer incidence. We re-analyzed the data from 43 meta-analyses to evaluate the associations between use of statins and cancer incidence. By only using a random-effects Although there was a weak or non-significant preventive effect of statin use on most cancer types, there was a suggestive level of evidence regarding the preventive effects of statin use on four cancer types . Re-analysis of association between statin use and leukemia incidence was performed with one eligible meta-analysis  consistip-value, an effect size with 95% CI [Although most of the re-analyses showed weak or non-significant evidence, the conventional interpretation of current meta-analysis is that there was preventive effect of statin use on cancer incidence in some cancer types, based on a random effects h 95% CI . AccordiIn addition, while most of the statistically significant individual meta-analyses showed that statins have a preventive effect on cancer, one meta-analysis of observational studies on association with statin and non-melanoma skin cancer suggested that there was a positive relationship between statin and non-melanoma skin cancer . Yang etp-value of the meta-analysis, between-study heterogeneity, small study effect, and 95% PI for the grading the level of evidence, which is more related to quantitative values [Our study evaluates the strength of evidence using multiple values presented or calculated in each meta-analysis. The strength of evidence reinforces the results from the meta-analyses and helps choose the best evidence. Various methods for assessing the evidence level are presented, yet there is no definite grading method for an umbrella review ,68. Recee values ,21,69.In addition, substantial heterogeneity is an issue in systematic review and meta-analysis. It is essential to explain and manage the heterogeneity to underline the validity of the respective findings . Umbrellevidence ,72. HoweIn the eligible meta-analyses, overlapping meta-analyses on the same topic were frequently reported . OverlapResults of the meta-analysis can be influenced by study design. Aromataris et al. reported that the types of studies should be matched in systematic reviews, and meta-analyses to be considered for its primary objective . In our p-value could not be assessed if there were only two or fewer individual studies. There were also some missing data in the largest study effect when there was no number of population data in individual studies. Finally, 95% PI, between-study heterogeneity, and publication bias may not be definitive criteria for assessing the strength of the evidence.Our study has several limitations. First, we only assessed individual studies from systematic reviews and meta-analyses eligible for re-analysis, and, therefore, some very recent individual studies might have been missed. However, considering that even a very recent updated meta-analysis for one cancer missed many individual studies, even though thorough search strategy was performed using many search sites such as PubMed, Embase, Scopus, Cochrane database, etc., we think that one should also check the individual studies from previous meta-analyses when updating the meta-analysis. Second, individual studies can have biases, but assessing the quality of individual studies was beyond the scope of our review. Third, exploring the association between dose and types of statins and cancer incidence was also beyond the scope. Likewise, due to a lack of applicable data, we could not stratify the effect of statins by participant age or duration of treatment, which may be the parameter needed to evaluate the true association. Fourth, there were statistical limitations. A 95% PI and Egger Nevertheless, in summary, we extensively re-analyzed meta-analyses on the associations between statin use and cancer incidence. In 10 of 18 studies there were significant relationships between statin use and cancer incidence. Although many meta-analyses of RCTs and observational studies reported significant associations between statin use and cancer incidence, only a small portion of these associations were without biases. Also, there was an individual meta-analysis reporting increased risk of cancer associated with statins use, which should be carefully interpreted by researchers and clinicians. Future studies should include more precise individual data, assessment of potential bias, and updated meta-analyses with more qualified RCTs and observational studies. We suggest that clinicians carefully consider the effects of statins on incidence of different types of cancer on the basis of the findings of our study.


Oral and pharynx cancer represent a serious global problem, reaching an incidence of half a million cases annually. The role of tobacco and alcohol have been studied and proven to be one of its risk factors. We also know that mouthwashes contain a variable percentage of alcohol, so there is a reasonable concern about their role in carcinogenesis. To answer the PICOS  question: "Do patients (Population) who use alcohol-based mouthwashes (Intervention) compared to those who do not use them (Comparison) have higher acetaldehyde levels in saliva or higher risk of oral cancer development? (Outcomes)" Meta-analyses, systematic reviews, randomized and non-randomized clinical trials, case-control studies, and prospective and retrospective cohort studies were included (Study). Two independent authors conducted literature screening through MEDLINE, Scopus and the Cochrane Library, and they also conducted article and data extraction to undertake quality analyses. The main outcome measures were salivary acetaldehyde levels or the risk of oral cancer development. The most relevant data was extracted and the risk of bias from the studies included was also evaluated. Out of 497 potentially eligible papers, 8 studies were included in the qualitative analysis which include a total of 43,499 subjects: two meta-analyses, a clinical trial, three case-control studies and two cohort studies. One study  found a relationship between alcohol mouthwash and oral cancer, two studies  found this relationship when a high frequency of mouthwash was present, three studies  failed to find this relationship and 2 studies (n = 58) found a temporary increase of acetaldehyde levels in saliva after alcohol mouthwash. It cannot be guaranteed that the use of mouthwash represents an independent risk factor for the development of head and neck cancer. However, the risk does increase when it occurs in association with other carcinogenic risk factors. Key words:Oral Cancer, Oropharyngeal Neoplasms, Mouthwashes, Mouth rinse, Ethanol, Acetaldehyde, alcohol. There is an estimated incidence of oral and pharynx cancer of half a million per year resulting in 250.000 deaths annually in United States.Most of them are squamous cell carcinoma of the oral cavity (SCC). The main risk factors for head and neck cancer are tobacco and alcohol; the risk increases when both factors are present. Approximately 70% of cancers in this region could be explained by exposure to one or both factors . Currently, poor oral hygiene and chronic infection from Human Papillomavirus (HPV) could be added to the main risk factors of oropharyngeal cancer, highlighted in young patients ,4.This risk from alcohol consumption increases ten times in heavy drinkers compared to abstainers or irregular drinkers . The totAlthough there is a short period of exposure time to alcohol, this seems to increase the permeability of the oral mucosa to potential carcinogens, as well as, specific Nitrosamines of tobacco .Many mouth rinses contain alcoholic concentrations between 5 - 27%. This is usually used as a solubilizer, stabilizer, preservative, anti-plaque efficacy enhancer and as a way to obtain a distinctive flavor. On the other hand, in a similar way to the alcoholic solutions ingested, mouthwashes with alcoholic concentrations between 18 - 27% also potentiate the effect of the essential oils to achieve a high penetration in soft tissues (in 30 seconds)  ,7. This The mechanism for the acetaldehyde production of ethanol can come from the metabolism of ethanol itself through oral microflora or conversion by the epithelial cells. It has been shown that poor oral hygiene increases the production of acetaldehyde, in addition to other factors (smoking and heavy drinking), producing a carcinogenic potential in the oral cavity and upper airways ,7.During the last thirty years, there has been an attempt to find a possible association between the use of mouthwash with alcohol and its relationship with oral cancer. However, epidemiologically, there has been no conclusive evidence. Few epidemiological studies are found in the literature and they have contradictory results .Many of the commercially available mouth rinses contain considerable amounts of ethanol. Most of the adult population use this type of mouth rinse, so it could be transcendent in public health to certify its possible carcinogenic potential.et al.  . A meta-- Selection criteriaInclusion criteria were based on the following PICOS question: patients with no age, gender or medical condition restrictions (Patient); alcohol-based mouthwashes use (Intervention); no mouthwashes use (Comparison); oral cancer or the acetaldehyde levels in saliva (Outcome). Meta-analyses, systematic reviews, randomized and non-randomized clinical trials, case-control studies and prospective and retrospective cohort studies were included. Just articles published in the last 10 years were included and no language restrictions were applied.Animal studies, case series, case reports and technical notes were excluded. Studies of digestive tract cancer not including the oral area or based on alcohol sugars were also excluded.The main outcome variable was the level of acetaldehyde in saliva. The secondary outcome variable was the risk of oral cancer development.- Search strategiesAn electronic search in Pubmed (MEDLINE), Scopus and the Cochrane Library  databases was conducted until April 13th, 2018. Even though the search was initially limited to the last 10 years, the first studies began in 2006, so in the end, articles published between 2006 and April 13th, 2018 were included.The search strategy was:  OR ) OR ) OR ) OR ) OR )) OR .- Study selection and data selection processAfter eliminating duplicate studies, two independent examiners  selected the studies by first screening the titles and abstracts of the articles to determine their eligibility. The full text of the selected articles were then evaluated according to selection criteria. Any discrepancies were resolved by a third reviewer (CGE). The excluded articles and the reasons for their exclusion were recorded in this final stage Fig. . Two Coh- Data extractionThe data was extracted from the articles selected by one of the authors (MUB). After analyzing all the selected articles, the following data were identified: 1) the author or authors, 2) year of publication, 3) country of origin, 4) design of the study, 6) details of the participants, 7) intervention, 8) comparison, 9) outcomes and 10) follow-up. If the included studies had a homogeneous methodology, a meta-analysis of the data would be possible.- Risk of bias in individual studiesAs part of the data extraction process, two reviewers  independently assessed the risk of bias in the included studies.For cohorts and case-control studies, the risk of bias was assessed according to the Newcastle-Ottawa Scale (NOS)  Table 1Table 1. For clinical trial or meta-analysis studies, the quality evaluation system was SIGN , a categ- Study selection and descriptionThe flow chart for the selected articles used in this systematic review can be seen in Fig. 18 articles were excluded according to the selection criteria. The exclusion reasons were: literary reviews -18, lettFinally, 8 studies were included in the qualitative analysis  5,28-34-3428-34.- Risk of bias assessmentAs shown in Both clinical trial  and one - Data extraction: qualitative synthesisAmong all the studies, a total of 43,499 subjects were included. In one of the meta-analysis   the numb) of the other included articles in the present study, 21,754 were male and 7,469 were female. The age ranged between 22 and 75 years old. The descriptive variables of the included studies can be found in et al.  also had an increased risk of upper airway cancer , as well as those using it more than once per day .et al. (p<0.0001) and although it is a positive OR, there was no significant difference if the use of mouthwash was once a day  or less than 1 use/day . All this data was obtained in centers located in Latin America . However, in both cases no differences were found between the group that brushes their teeth ≥2 times/day and the group that never brushes it . These results were obtained when patients were analyzed without considering possible confounding factors such as smoking and alcohol. The tobacco and alcohol habit modified the risk of cancer according to the use of mouthwash ; no significant differences were found between subjects with oral cancer and poor oral health (p<0.01), but no differences were found between occasional tooth washing and 1-2 washes/day (p=0.274).Kocaelli et al.  studied l health . The aceshes/day . Finallyet al. , 1.42 (95%CI = 0.91 – 2.24) and 1.7 (95%CI = 0.86 – 3.35), respectively, with I2 = 76% and Chy-square - Analysis according to smoking or alcohol habitsp=0.06) with no statistically significant differences. On the other hand, in ex-smoker patients an OR: 4.98  and in smokers an OR: 9.15 , both with statistically significant differences. The differences were not significant when the use of mouthwash was 1 time/day ; OR: 1.94 ; OR: 1.53  respectively), obtaining a positive dose-dependent trend  and OR: 1.25  respectively). Smokers with poor oral hygiene obtained an increased risk for oral, pharynx and larynx cancer compared to those smokers with good oral hygiene, with an OR: 3.60   .p=0.23) and OR: 1.20  respectively; however they were observed in smokers   .et al. . However, no significant differences were found between heavy smokers and moderate smokers . Oral hygiene was not assessed in the European case due to a limited sample. In the American sample, there were no significant differences in drinkers who presented good or bad oral hygiene , but they were significant in drinkers, with a higher risk for subjects with poor oral hygiene than those with good oral hygiene . In this case, there were no significant differences between the number of tooth washes per day and the consumption of alcohol in terms of the risk of oral, pharynx or larynx cancer . On the contrary, there seemed to be a non-statistically significant predilection in those subjects who used 1 or 2 mouthwashes, with a dose-dependent relationship. Gandini et al. , it is necessary to assess whether a high frequency of rinses may increase the risk of oral cancer. Several classic studies indicate that the use of this mouthwash significantly increases the risk of oral cancer . Guha etet al.  and Ahre (et al.  obtainedi et al. . Ahrens  (et al.  and Boff (et al.  observed (et al.  argued t (et al. ,29,33.In patients without other risk factors, the carcinogenic effect seems to be low and still arguable -32. On tTherefore, it seems that the evidence is limited to the carcinogenic effect of the mouthwash use in patients without other risk factors for oral, oropharynx or head and neck cancer. There might be a trend which indicates the presence of this effect but its existence is still not clear in a meaningful way, and if such an effect existed it would be low.However, although there were limited results in the included studies due to the heterogeneity between them and the subjects ,34, it iIn conclusion, alcohol-based mouthwash consumption significantly increases salivary acetaldehyde levels in the first few minutes. However, no evidence exists if long-term salivary acetaldehyde levels may increase with a high frequency of mouthwash use. There is still insufficient evidence of whether the use of alcohol-based mouthwash is an independent risk factor for oral or oropharynx cancer. Nonetheless, it does increase the risk when it occurs concomitantly with other risk factors such as smoking or alcohol.


P > 0.05), advanced CRA (P > 0.05) and multiple CRA (P > 0.05). In conclusion, our findings support the hypothesis that red and processed meat intake was associated with an increased CRA incidence but not for CRA recurrence.The associations between red and processed meat intake and colorectal adenoma (CRA) incidence and recurrence are inconclusive. We performed a systematic review and meta-analysis to analysis these associations. We conducted a systematic search of PubMed, EMBASE and Web of Science up to December 2016. The relative risks (RRs) and 95% confidence intervals (CIs) were assessed. Subgroup analyses, dose-response-analyses, subtype analyses and analyses of CRA locations were also conducted. Twenty-seven studies that involved 208,117 participants and 19,150 cases met criteria. The RRs of the highest versus lowest intakes for CRA incidence were 1.23 (1.15–1.31) for red meat and 1.15 (1.07–1.24) for processed meat. Dose-response analyses for meat per 100 g/day yielded the results were consistent with the original analyses, with 1.14 (1.07–1.20) for red meat and 1.27 (1.03–1.50) for processed meat. Additionally, there were no associations between red and processed meat intake and CRA recurrence, including total CRA ( According to the Cancer Statistics 2017, colorectal cancer (CRC) is the third most frequently diagnosed cancer, with 135,430 estimated new cases and 50,260 estimated deaths in 2017 occur in the United States  on the aTwenty-seven studies met the criteria and provided 34 separate estimates  for CRA incidence, and 20 separate estimates  for CRA recurrence   for sample size, publication year and all adjustments   with low heterogeneity (y drugs) .We further examined the associations between red meat intake and the CRA location. Ten studies were included and the analyses suggested significantly different results, with positive results for distal colon adenoma  and negative results for proximal colon adenoma  and rectal adenoma  (Table P < 0.01). Furthermore, we checked for nonlinearity of the dose-response relationship and the evidence showed that the best-fitting model was nonlinear model (Pnonlinearity < 0.01)  suggested that the CRA incidence increases by 14% for each 100 g/day increase in red meat intake ( < 0.01) .P = 0.94) did not suggest significant evidence of publication bias. The sensitivity analyses of the highest vs lowest categories showed that the changes in the recalculated RRs were not significant, with a range from 1.19 (1.11–1.28) when excluding Fu 2011 [The funnel plot  and Egge Fu 2011  (17.1%)  Fu 2011  (17.8%).P = 0.05, I2 = 52%)  with heterogeneity , 0.99 (0.82–1.20) for advanced CRA without heterogeneity  and 0.93 (0.75–1.14) for multiple CRA with low heterogeneity   for total CRA without heterogeneity (I2 = 0%) , Table 3P = 0.03, I2 = 66%)  with heterogeneity   for sample size, publication year and all adjustments  excluded dietary fiber intake and family history of CRC/polyps  with low heterogeneity  and negative results for rectal adenoma  (Table P = 0.03). Furthermore, we checked for nonlinearity of the dose-response relationship and the evidence showed that the best-fitting model was nonlinear model (Pnonlinearity< 0.01)  suggested that the CRA incidence increases by 27% for each 100 g/day increase in processed meat intake (y< 0.01) .P = 0.77) did not suggest significant evidence of publication bias. Notably, the sensitivity analyses of the highest vs lowest categories showed that the changes in the recalculated RRs were significant, with a range from 1.12 (1.03–1.22) when excluding Fu 2011 [The funnel plot  and Egge Fu 2011  (20.5%) P = 0.73, I2 = 0%)  without heterogeneity , 1.14 (0.95–1.37) for advanced CRA with low heterogeneity  and 1.09 (0.73–1.62) for multiple CRA with significant heterogeneity  (Four studies were included in the comparison of the highest vs lowest categories when further stratified by CRA type. The results were 1.10 0.94–1.30) for total CRA with low heterogeneity (2 = 69%) , Table 30 for totOn the one hand, our findings supported the hypothesis that high intakes of red meat and processed meat increased the CRA incidence. Similarly, the dose-response analyses found positive associations for red meat and processed meat. Furthermore, the results of subgroup analyses that were based on the main adjustment for confounders were consistent for each confounder and similar to the original analyses. Additionally, subtype of analyses for red meat (beef) and processed meat (bacon) yielded the consistent results with the original estimates. We also performed the analyses of CRA locations, which further showed that positive associations were observed in distal CRA for red meat and in proximal CRA for processed meat. We specifically analyzed the ratio of red meat/white meat, and the positive results indicated that the types of meat and the ratio may be associated with CRA risk.On the other hand, we also examined the associations between red and processed meat intake and CRA recurrence; the analyses indicated that red meat and processed meat intake was not associated with the recurrence of total CRA, advanced CRA and multiple CRA. Overall, our findings highlight the associations between red and processed meat intake and CRA risk, which may be a reference to update the dietary recommendations.Several potential mechanisms may contribute to the effects. First, the positive associations between red and processed meat intake and CRA risk may be biologically plausible. Cooking red and processed meat is considered one of the major sources of carcinogens, such as heterocyclic amines (HCAs), polycyclic aromatic hydrocarbons (PAHs), nitrate and N-nitroso compounds (NOCs), which are believed to play important roles in the etiology of cancer –39 and aThere are several limitations in this meta-analysis. First, information on several of the major confounders, such as the intake of vegetable and fruit, could not be provided in all studies. Thus, the findings should be considered carefully due to possible confounding. Second, the different exposure ranges from the lowest to highest categories among included studies contributed to possible heterogeneity. Nevertheless, we adopted the RRs for the comparison of the highest to lowest categories. Additionally, dose-response analyses were conducted to verify the estimate. Third, the cooking methods, storage conditions, production methods and nutrient contents of meat may differ among studies, and the measurement errors to assess meat intake may lead to bias. We cannot thoroughly exclude the potential residual confounding. Finally, the language of studies was limited to English, and several studies with null estimates might not have been reported. Thus, we detected publication bias using the funnel plot, Egger's test, and the sensitivity analysis, which suggested the negligible publication bias.Our analysis has several strengths. First, this study provided sufficient robust, reliable and current evidence and increased the statistical power based on a substantial sample size and a quantitative synthesis of the eligible data. These data Second, we examined the association between red and processed meat intake and CRA incidence  and recurrence . We performed subtype analyses of white meat (poultry and fish) to further explore the association. In addition, we conducted subgroup analyses for CRA according to the main risk factors  and the main confounding factors between studies  to explore the stability of pooled estimates. Third, dose-response analyses were performed to further assess the association rather than simply conducting categorical comparisons. All the independent analyses provided detailed data and increased the statistical power and the strength of our conclusion. Fourth, the study selection and data extraction were performed independently and in duplicate by two authors, which increased the validity of our findings. Finally, the heterogeneity and publication bias of this meta-analysis was negligible, which increased the reliability of our results.We systematically searched PubMed, EMBASE and Web of Science for studies up to December 2016 using the following search terms: “meats, meat, beef, pork, mutton, veal, lamb, horse, bacon, ham, salami, sausage, hot dogs, lifestyle, food, foods, diet and dietary” in combination with “neoplasm, neoplasms, neoplasia, adenoma, adenomas, cancer, cancers, adenocarcinoma, polyp and polyps”. The two sets were combined individually, and two authors (ZZ and ZY) independently judged the eligibility criteria. Additionally, the reference lists of studies were searched manually to identify eligible literature.Selection criteria were as follows: studies that diagnosed patients with endoscopy by histological features and biopsy that were consistent with the diagnostic gold standard were included; data that could not be combined were excluded; data that were incomplete were excluded; studies published as original articles were included; pooled analyses, systematic reviews, meta-analyses, narrative reviews, editorials, case reports, letters and comments were excluded; colorectal adenocarcinoma, precancerous lesions and other colorectal tumors were excluded; the included studies were limited to those involving humans and the language was limited to English.Two authors (ZZ and ZY) assessed the quality of included studies independently, and discrepancies in interpretation were resolved by a consensus decision made by the third author (QZ). Study quality was assessed using the Newcastle-Ottawa Scale (NOS) for observational studies  and the The STATA version 12.1  and RevMan5.3  were used for data synthesis and analysis.A random-effects model was used if there was heterogeneity among studies, and a fixed-effects model was used without heterogeneity. The median or mean level for each category was assigned to each corresponding RR. The non-linear dose-response analysis was conducted using the method described by Greenland et al . The stuI2 statistic  [P < 0.1 was considered representative of significant heterogeneity) were used to detect the heterogeneity among studies. Subgroup analyses were conducted to explore the sources of heterogeneity by study design, publication year, geographic area, sample size and adjustments .The geneity)  and the P < 0.1 of Egger's test was considered significant publication bias. The sensitivity analysis was performed to investigate the influence of each study on the pooled risk estimate by removing one study in turn.Publication bias was evaluated using the funnel plot, Egger's test  and a seThe present analysis provided evidence that the intake of red meat and processed meat was associated with an increased incidence of CRA. No associations were found between red meat and processed meat intakes and CRA recurrence.


Metabolic Syndrome (MetS) constitutes a relevant public health burden. Several studies have demonstrated the association between diet and MetS. We performed a systematic review and meta-analysis to provide an estimate of the association between dietary patterns defined through a posteriori methods and MetS. A literature search on PubMed, Web of Science, and Scopus databases, up to March 2019, was conducted to identify all eligible case-control, prospective, or cross-sectional studies involving adult subjects of both sexes. Random-effects models were used. Heterogeneity and publication bias were evaluated. Stratified analyses were conducted on study characteristics. Forty observational studies were included in the meta-analysis, which identified the “Healthy” and the “Meat/Western” dietary patterns. The “Healthy” pattern was associated with reduced MetS risk : 0.79–0.91) and significantly decreased the risk in both sexes and in Eastern countries, particularly in Asia. Adherence to the “Meat/Western” pattern increased MetS risk  and the association persisted in the stratified analysis by geographic area  and study design. Lifestyle is linked to risk of developing MetS. The “Healthy” and “Meat/Western” patterns are significantly associated with reduced and increased MetS risk, respectively. Nutrition represents an important modifiable factor affecting MetS risk. Metabolic Syndrome (MetS) has become a relevant public health concern  because MetS is a pathophysiological state and a cluster of interrelated factors including abdominal obesity, insulin resistance, dysglycemia, hypertension, and dyslipidemia (triglycerides and HDL-C—high-density lipoprotein cholesterol) . The diaAccording to literature, the consumption of specific foods or nutrients is strongly related to the risk of developing MetS ,7,8,9. NThe aim of our systematic review and meta-analysis is to investigate and provide an estimate of the association between dietary patterns defined by a posteriori methods and MetS risk in adults.The present meta-analysis was conducted following the MOOSE  guidelines  and PRIShttp://www.ncbi.nlm.nih.gov/pubmed/), Web of Science (http://wokinfo.com/) and Scopus (https://www.scopus.com/) databases to identify all the original articles on the association between dietary patterns and MetS. The literature search included the following search medical subject headings (MeSH) and key words: (“Metabolic Syndrome” OR MetS) AND  AND .We conducted a comprehensive literature search, without restrictions, up to 31 March 2019 through PubMed  evaluated the relationship between dietary patterns derived by a posteriori methods, such as principal component analysis (PCA), factor analysis (FA), and principal component factor analysis (PCFA), and by reduced rank regression  and MetS in adults; (ii) used a case-control, prospective or cross-sectional study design; (iii) reported odds ratio (OR), relative risk (RR) or hazard ratio (HR) estimates with 95% confidence intervals (CIs). For each potentially included study, two investigators independently carried out the selection, data abstraction, and quality assessment. Disagreements were resolved by discussion or in consultation with the third author. Although useful to have background information, reviews and meta-analysis were excluded. No studies were excluded for weakness of design or data quality.p-value for trend, and confounding factors adjustment. When multiple estimates were reported in the article, we pulled out those adjusted for the most confounding factors.For each selected study, we extracted the following information: first author’s last name, year of publication, country, study design, sample size , population characteristics , duration of follow-up for cohort studies, MetS assessment method, dietary assessment and dietary pattern identification methods , characteristics of the dietary assessment method, name given to the dietary patterns and their characteristics, cutoff points of the different categories of adherence to the dietary pattern , risk estimates with 95% CIs for the different categories of adherence, The estimated overall effect-size statistic was the average of the logarithm of the observed OR  associated with the highest versus the lowest level of adherence to the different dietary patterns. We used the random-effects model to calculate the summary OR and 95% CIs. We restricted the analysis to the dietary patterns defined a posteriori. Since the labeling of the patterns is arbitrary and the dietary patterns are population-specific, we considered only those patterns sharing most foods with similar factor loadings. For the inclusion in the meta-analysis, the two most common dietary patterns with similar factor loading of principle components were identified in 38 studies out of 40)  40,41,4,56,57,58,57,5856,The second dietary pattern, named “Meat/Western”, had a high loading of red meat, processed meat, animal fat, eggs and sweets. The included articles labeled this pattern as “Western” ,53,54,582 statistic were used to evaluate heterogeneity in results across studies [2 statistic yields results ranged from 0% to 100%  [p-value of <0.05. The analyses were performed using the ProMeta Version 3.0 statistical program .The chi-square-based Cochran’s Q statistic and the I studies . The I2 geneity) . Resultsgeneity)  and Eggegeneity)  to detecn = 90), Web of Science (n = 227) and Scopus (n = 143) databases identified a total of 460 articles. Duplicates (n = 158) were removed and the remaining 302 records were identified for title and abstract revision  as reference; one study was carried out on transplant recipients; and one study was carried out on type 2 diabetes. Therefore, at the end of the selection process, 40 studies were enclosed for the identification of the different dietary patterns in the systematic review and meta-analysis ,56,57,58General characteristics of the 40 studies evaluating the association between adherence to a posteriori dietary patterns with MetS risk are shown in These studies were published between 2007 and 2019. Eight studies were conducted in Korea ,40,43,45Thirty-one studies used a food frequency questionnaire  ,56,57,58We identified two common dietary patterns with similar factor loading of principle components: “Healthy” and “Meat/Western” patterns. Thirty-eight out of 40 articles included in the systematic review were used for the overall risk estimation. Two studies ,48 were The overall analysis showed that the MetS risk significantly decreased in association with the adherence to the “Healthy” pattern  and significantly increased in association with the adherence to the “Meat/Western” pattern . These results did not essentially change when the studies ,33,52 noIn the “Healthy” pattern meta-analysis, the stratification by study design showed a significant reduced MetS risk in the cross-sectional studies only . StratifIn the “Meat/Western” pattern meta-analysis, the stratification by study design showed a significantly higher MetS risk in both cohort and cross-sectional studies . SimilarThe high heterogeneity in the pooled analysis of both “Healthy” and “Meat/Western” patterns was slightly reduced in the stratification by geographic area.Sensitivity analyses suggested that the estimates were not substantially modified by any single study. Small changes were found in the risk estimates after removal of the outlier studies by Naja et al.  (OR = 0.p = 0.005) and in cross-sectional studies (p = 0.016), but not by the Begg’s method (p = 0.011) (p = 0.021) and by the Begg’s test in men (p = 0.042)  . In the = 0.042) .The funnel plots of the meta-analyses on the “Healthy” pattern and on the “Meat/Western” pattern are shown in Our systematic review and meta-analysis investigated the effect of dietary patterns extracted via a posteriori methods on MetS risk. According to literature, several different health outcomes are associated with unhealthy and healthy dietary patterns. In particular, the Western/unhealthy pattern increases the risk of cancer in different sites ,66,67,68Considering the 40 included articles, we identified two prevalent dietary patterns: “Healthy” and “Meat/Western”. The “Healthy” pattern was associated with a lower MetS risk and significantly decreased the risk in both sexes and in Eastern countries, particularly in Asia. Adherence to the “Meat/Western” pattern was positively associated with MetS risk and this association persisted in the stratified analysis by geographic area and study design. Similarly, the recent meta-analyses by Shab–Bidar et al.  and RodrAccording to our findings, the “Meat/Western” pattern significantly increased MetS risk of 20% in Asia, 15% in Europe and 33% in America. In dietary patterns derived a posteriori, the factor loadings indicate the most commonly consumed foods, reflecting the cultural influence on food consumption ,75. It iDespite the influence of sex-related factors on MetS , we obseThe “Meat/Western” pattern, characterized by high intake or red and processed meat, eggs, refined grains, and sweets, resulted associated with an increased (+19%) MetS risk. These foods plausibly represent the main cause of the observed effect on MetS risk, particularly meat ,82, sincOur results showed the association of the “Healthy” pattern with a lower (−15%) MetS risk. The healthy patterns are characterized by the consumption of foods with high content of vitamins, minerals, antioxidants, fiber, MUFA and n-3 fatty acids, which could contribute to explain the protective effect of the “Healthy” pattern on MetS. Indeed, higher adherence to healthy dietary patterns is associated with a lower risk of glucose intolerance, weight gain, inflammation, insulin resistance and a higher level of HDL cholesterol .post hoc event. Only the OR of the highest and the lowest quantile of healthy or unhealthy dietary patterns were included in our analysis, limiting the evaluation of the presence of any trend. Finally, some studies reported risk estimates for quintiles, others for quartiles, and others for tertiles. As dietary intakes are influenced by sex, race/ethnicity, and societal factors, our findings should be considered in the different geographic contexts. Thus, these aspects may have affected the reproducibility of the association between dietary patterns and MetS.The main limitation of our study is that the risk of developing MetS could be associated with dietary patterns other than the two  discussed in this meta-analysis. Differences in the populations in study and in the referral values for MetS diagnosis represent another study limitation and result in heterogeneity. Indeed, the high heterogeneity may be related to the wide variability in dietary data collection and analysis, in the various and not uniformly adjusted confounding factors, and in the identification of the dietary patterns. Heterogeneity is more evident in the analysis on “Meat/Western” pattern, as a possible consequence of the difficulty in characterizing this pattern across the selected studies. Another limitation is that pooled data were directly driven by the included studies, presenting their own weaknesses in study design. Moreover, the cross-sectional nature of many included studies precludes causal inference and the dietary pattern may represent a To further advance this field of research, future studies are needed to examine the association between dietary patterns in geographic context not yet described and MetS, and to evaluate the impact of dietary patterns on the determinants of MetS.A protective effect on MetS is attributed to adherence to the “Healthy” pattern, which is characterized by high consumption of fruit, vegetables, whole grains, poultry, fish, nuts, legumes, and low-fat dairy products, whereas the “Meat/Western” pattern is positively associated with MetS. Nutrition is one of the most important modifiable factors affecting health. Public health efforts should aim to adopt healthy dietary patterns and to reduce the burden of MetS, providing guidance for nutritional intervention. For further advance in research, more prospective studies are needed to investigate the association between dietary patterns and MetS in each gender and in different geographic context.


Scarce literature has investigated the association of potential functional genetic variants of IRAKs with Hepatitis B Virus- (HBV-) related hepatocellular carcinoma (HCC). Interleukin 1 receptor associated kinases A case-control study with 1,538 HBV-positive HCC patients and 1,465 chronic HBV carriers was conducted to evaluate the effects of common missense variants of IRAK family members on HCC. Proliferation assays and real-time polymerase chain reactions were carried out to evaluate the functions. Multivariable adjusted logistic regression was adopted to estimate effect size and identify risk factors. IRAK4 (p.Ala428Thr) was positively associated with HBV-related HCC risk . Functional annotation indicated that rs4251545 reduced its own expression in liver (P = 0.031). Further molecular functional analysis detected that rs4251545 increased the proliferation rate of L02 cells (P < 0.05). Meanwhile, rs4251545 reduced mRNA expressions of IL-6, IL-8, CXCL-1, and CXCL-2 in L02 cells (P < 0.01). Association analysis indicated that rs4251545 A allele of rs4251545 of IRAK4 (p.Ala428Thr) modified the susceptibility to HBV-related HCC via increased proliferation rate and reduced production of inflammatory cytokines and chemokines. Further well-designed experiments are warranted to validate our findings. Globally, HCC was the second leading cause of cancer-related death in men . The incκB or IRF was activated and released the proinflammatory cytokines, such as IL-6 and IL-8, and interferons  and this difference was attenuated in the Eastern China population (Shanghai). Overall, the whole subjects did not present heterogeneity in terms of geographic locations , significantly increased the risk of HCC [(OR = 1.30 (1.09–1.54), P < 0.05). However, no statistical difference was observed in HepG2.2.15 and HepG2 cells , while no significant impact was detected in HepG2 cells  and chemokines (CXCL-1 and CXCL-2). These results indicated that IRAK4 should play an important role in the pathogenesis of HCC, which could be set as a target for the future individual treatment.Present study used a case-control design and found an unfavorable association between rs4251545 (IRAK4 p.Ala428Thr) and HBV-related HCC even after adjusting basic social demographic factors, which could be attributed to the diminished signaling activation of NF-Our results were consistent with a previous study which found that rs4251545 was associated with 1.68- to 4.99-fold increases in the risk of developing breast cancer in African-American women . MoreoveκB signaling and IFN-α and correlated with reduced HCV auto clearance rate and increased risk of chronic HCV infection [The plausibility of reduced proinflammatory and increased proliferation rate of cells could be attributed to the unbalanced immunity reaction. Several studies found that dysregulation of TLR signaling pathway activation might shift the balance between the production of pro- and anti-inflammatory cytokines and would have a profound effect on the risk of infection, chronic inflammation, and cancer . Our prenfection . Anothernfection . InteresSeveral limitations in present study should be mentioned: first, several relevant factors that might partially explain the function of immunity at different stage of carcinogenesis, such as cirrhosis status and viral load, were not included into the analysis due to data availability. Second, external population-based validation study should further solidify our results. Third, the population only restricted to Chinese population which would limit the ability of extrapolation. Finally, further comprehensive functional analyses, such as transwell assay, ELISA assay coimmunoprecipitation, or EMSA, should be carried out to deepen our understanding of the molecular mechanism of the pathogenesis in HCC. Overall, it is the first attempt to evaluate the potential functional variants of IRAKs in the pathogenesis of HCC with a moderate sample size. Our study further confirmed that IRAK4 is involved in progression of HBV-related HCC. However, more well-designed population validation and molecular assays were warranted in the future analysis.