what causes the body to release c-reactive protein into the bloodstream?

• Periodical Listing
• Arch Med Sci
• five.xiv(4); 2018 Jun
• PMC6040119

Arch Med Sci. 2018 Jun; 14(4): 707–716.

Upshot of magnesium supplements on serum C-reactive protein: a systematic review and meta-analysis

Mohsen Mazidi

^oneFundamental State Laboratory of Molecular Developmental Biology, Institute of Genetics and Developmental Biology, Chinese Academy of Sciences, Chaoyang, Beijing, Communist china

²Institute of Genetics and Developmental Biology, International Higher, Academy of Chinese University of Scientific discipline (IC-UCAS), West Beichen Road, Chaoyang, Prc

Peyman Rezaie

³Biochemistry and Nutrition Research Centre, School of Medicine, Mashhad University of Medical Science, Mashhad, Islamic republic of iran

Maciej Banach

^ivDepartment of Hypertension, Chair of Nephrology and Hypertension, Medical Academy of Lodz, Poland

⁵Smoothen Mother'southward Memorial Hospital Research Institute (PMMHRI), Lodz, Poland

⁶Cardiovascular Enquiry Middle, University of Zielona Gora, Zielona Gora, Poland

Received 2017 Aug 10; Accustomed 2017 Aug 18.

Abstract

Introduction

The aim of the study was to undertake a systematic review and meta-assay of prospective studies to determine the consequence of magnesium (Mg) supplementation on C-reactive protein (CRP). Design: Systematic review and meta-analysis of randomised controlled trials (RCTs).

Textile and methods

Data sources: PubMed-Medline, Web of Science, Cochrane Database, and Google Scholar databases were searched (up until December 2016). Eligibility criteria: Randomized controlled trials evaluating the impact of Mg supplementation on CRP. We used random furnishings models meta-assay for quantitative data synthesis. For sensitivity analysis was used the leave-one-out method. Heterogeneity was quantitatively assessed using the I ² index. Main outcome: Level of CRP later Mg supplementation.

Results

From a total of 96 entries identified via searches, viii studies were included in the final selection. The meta-analysis indicated a meaning reduction in serum CRP concentrations following Mg supplementation (weighted mean difference (WMD) –ane.33 mg/50; 95% CI: –2.63 to –0.02, heterogeneity p < 0.123; I ⁱⁱ = 29.i%). The WMD for interleukin vi was –0.xvi pg/dl (95% CI: –three.52 to 3.26, heterogeneity p = 0.802; I ² = 2.3%), and 0.61 mg/dl (95% CI: –2.72 to ane.48, p = 0.182, heterogeneity p = 0.742; I ⁱⁱ = half-dozen.i%) for fasting blood glucose. These findings were robust in sensitivity analyses. Random-furnishings meta-regression revealed that changes in serum CRP levels were independent of the dosage of Mg supplementation (slope: –0.004; 95% CI: –0.03, 0.02; p = 0.720) or duration of follow-up (gradient: –0.06; 95% CI: –0.37, 0.24; p = 0.681).

Conclusions

This meta-analysis suggests that Mg supplementation significantly reduces serum CRP level. RCTs with a larger sample size and a longer follow-upwardly menses should be considered for future investigations to requite an unequivocal reply.

Keywords: meta-analysis, magnesium, C-reactive protein

Introduction

Magnesium (Mg), every bit i of the near abundant minerals in the body, is essential for good health. The main nutrient sources of Mg are whole grains, legumes, nuts, and green leafy vegetables [1]. More recent evidence indicates that dietary intake of Mg has an impact on several metabolic and inflammatory disorders including hypertension [2], type 2 diabetes [3], metabolic syndrome [4], insulin resistance [five] and cardiovascular diseases [1, six]. Magnesium deficiency, either from inadequate intake, excess excretion or contradistinct homeostasis, is frequently suspected to be associated with the initiation of many symptoms and diseases [7]. A growing body of evidence supports the of import office of magnesium deficiency in the synthesis and release of pro-inflammatory cytokines and astute phase functions, in the impairment of peripheral insulin action, and in the progress of glucose metabolism disturbances [viii, ix]. It has been proposed that the anti-inflammatory response of Mg may contribute to the beneficial furnishings on reducing the levels of C-reactive protein (CRP) – a well-known indicator of acute or chronic inflammation [10]. In this regard, one study has reported an inverse human relationship between dietary magnesium intake and CRP levels in non-diabetic non-hypertensive obese individuals [eleven]. However, another cross-sectional study did not find an clan between dietary Mg intake and CRP levels [12]. While these findings suggest that magnesium plays a role in the inflammatory response, single studies to appointment have been express by sample size, research design and subject traits (gender, ethnicity, age, etc.), and underpowered to attain a comprehensive and reliable conclusion. On the other hand, dietary supplementation with Mg with different dosage and duration can have dissimilar effects on some indices of inflammatory and anti-inflammatory indexes. Meta-analysis has the benefit of overcoming this limitation by increasing the sample size. Hence, to better understand this issue, the nowadays study aimed to resolve this uncertainty past systematically reviewing the literature and performing a meta-analysis of all randomized control trials investigating the effects of Mg supplementation on serum CRP levels.

Material and methods

We followed the guidelines of the 2009 Preferred Reporting Items for Systematic Reviews and Meta-Analysis (PRISMA) statement [xiii, 14]. Due to the study design (meta-analysis) neither Institutional Review Board (IRB) approval nor patient informed consent was needed or obtained. The study protocol was registered with the International Prospective Register of Systematic Reviews, PROSPERO (registration no: CRD42016039482).

Literature search strategy

The principal exposure of interest was magnesium supplementation, while the primary event of interest was changes in serum C-reactive poly peptide levels subsequent to magnesium supplementation. We searched multiple databases including PubMed/Medline, Cochrane Central Register of Controlled Trials (CCTR), Cochrane Database of Systematic Reviews (CDSR), Spider web of Science and MEDLINE, until December 2016, using a combination of search terms available in Table I. We used the wild-card term '*' to increment the sensitivity of the search strategy. No linguistic communication brake was applied. This was complemented by hand searches of the reference list of eligible manufactures, and email correspondences with authors for additional data where relevant. All newspaper abstracts were screened by 2 reviewers (MM and PR) in an initial process to remove ineligible articles. The remaining manufactures were obtained in full text and assessed once more by the aforementioned 2 researchers who evaluated each article independently, carried out data extraction and quality assessment. Disagreements were resolved by discussion with a 3rd political party (HV).

Table I

Full search terms and strategy used for systematically reviewing the articles

No	Concept	Search terms
ane	Magnesium	(("magnesium"[Text Word]) OR "Mg"[Text Word])
2	C-reactive poly peptide	"high sensitivity C-reactive protein"[MeSH Terms] OR "high-sensitivity C-reactive poly peptide"[MeSH Terms] OR "C-reactive protein"[MeSH Terms] OR "high-sensitive C-reactive protein"[MeSH Terms] OR "high sensitive C-reactive protein"[MeSH Terms] OR "CRP"[Championship/Abstract] OR "hsCRP"[Title/Abstract]
3	Combination	one AND 2

Selection criteria

All prospective studies evaluating the effect of Mg supplementation on the outcomes of interest were included in this analysis. Eligible studies had to meet the following criteria: (1) prospective controlled trial with either parallel or crossover pattern and of patients treated with magnesium supplementation compared to a control grouping (either no Mg supplementation or placebo), (2) presentation of sufficient information on primary outcome at baseline and at the end of follow-up in each group or providing the net modify values. Exclusion criteria were: (i) non-clinical studies; (ii) observational studies with example–control, cantankerous-exclusive or cohort pattern; and (iii) studies that did not provide hateful (or median) plasma concentrations of our interested outcomes at baseline and/or at the end of trial. Narrative reviews, comments, opinion pieces, methodological, editorials, letters or any other publications defective main data and/or explicit method descriptions were also excluded. Study pick started with the removal of duplicates, followed by screening of titles and abstracts by two reviewers. To avoid bias, they were blinded to the names, qualifications or the institutional affiliations of the study authors. The agreement between the reviewers was excellent (κ index: 0.88; p < 0.001). Disagreements were resolved at a coming together between reviewers prior to selected manufactures existence retrieved (a catamenia chart is available in Figure i).

PRISMA flow chart for selection of studies

Information extraction and management

The total text of studies meeting the inclusion criteria was retrieved and screened to determine eligibility past ii reviewers (MM, PR). Following cess of methodological quality, the two reviewers extracted data using a purpose-designed data extraction form and independently summarized what they considered to exist the most important results from each written report. These summaries were compared and whatever differences of stance resolved by discussion and consultation with a third reviewer (HV). Additional necessary calculations on study data were conducted by the first reviewer and checked past the 2d reviewer. Descriptive data extracted included the first author'southward name, reference, yr of publication, land, blueprint, duration of the study, inclusion criteria, dose, age range, sample size, and male gender (%).

Quality assessment

A systematic assessment of bias in the included RCTs was performed using the Cochrane criteria [15]. We sued the following items for the assessment of each report: blinding of participants, allotment concealment, adequacy of random sequence generation, handling of drib-outs (incomplete result data), personnel and consequence assessment, selective issue reporting, likewise as whatever other potential sources of bias. According to the recommendations of the Cochrane Handbook, a judgment of 'yes' indicated low risk of bias, while 'no' indicated a high risk of bias. Labeling an item as 'unclear' indicated an unclear or unknown gamble of bias. Risk-of-bias cess was performed independently past 2 reviewers (MM and PR); disagreements were resolved by a third reviewer (HV).

Data synthesis

Following the Cochrane Handbook recommendations for calculating the effect size, nosotros used the mean (SD) change from baseline in the concentrations of the variables of interest for both command and intervention groups [sixteen]. In summary, nosotros calculated the internet changes in measurements (alter scores) as follows: measure at the end of follow-up – measure out at baseline. For RCTs, alter scores were calculated as (measure at the finish of follow-upwards in the treatment group – measure at baseline in the treatment grouping) – (measure at the finish of follow-up in the control grouping – measure out at baseline in the control group). In situations where just a standard error of the mean (SEM) was available, we estimated standard divergence (SD) using the following formula: SD = SEM × foursquare root (n), where n is the number of subjects [17]. In situations where the event measures were reported in median and range (or 95% confidence interval (CI)), nosotros estimated the mean (SD) values using the method described by Mazidi et al. [18]. When the issue variable was available only in the graphic form, the software GetData Graph Digitizer 2.24 [19] was used to digitize and extract the data. Claret glucose level was collated in mmol/fifty; a multiplication factor of 0.0555 was used to convert glucose levels from mg/dl to mmol/l as appropriate [20].

A random effects model (using the DerSimonian-Laird method) and the generic changed variance method were used to compensate for the heterogeneity of studies in terms of demographic characteristics of populations being studied and besides differences in study design and type of BAS existence studied [21]. Heterogeneity was quantitatively assessed using the I ² index. The I ² values < 50% and ≥ fifty% corresponded with the use of stock-still-effects and random-effects models, respectively. Nosotros expressed the effect sizes as the weighted mean departure (WMD) and 95% confidence interval (CI). To assess the influence of each study on the overall consequence size, we conducted a sensitivity analysis using the leave-one-out method (removing one study each time and repeating the assay) [22–24].

Meta-regression

Random-effects meta-regression was performed using the unrestricted maximum likelihood method to evaluate the clan between calculated WMD and potential moderator including dose of magnesium supplementation.

Publication bias

Potential publication bias was explored using visual inspection of Begg's funnel plot asymmetry, Begg's rank correlation and Egger's weighted regression tests. Duval & Tweedie 'trim and fill' and 'fail-safe N' methods were used to suit the analysis for the effects of publication bias [25]. Meta-analysis was conducted using Comprehensive Meta-Analysis (CMA) V3 software (Biostat, NJ) [26].

Results

Summary of searches and written report option procedure

A full of 96 unique citations were identified from searches, of which 72 records remained after removing duplicates. After screening via titles and abstracts, 15 manufactures remained for further evaluation, of which vii were excluded for the following reasons: not-human studies, genetic, or molecular studies (n = four); reviews or editorial articles (northward = 2) and not RCTs (n = one) (Figure 1). Therefore, viii studies with 349 participants were finally included in the meta-analysis.

Risk of bias cess

Several of the included studies were characterized by a lack of information about allocation concealment (due north = 1) or blinding of outcome cess (n = 2). Nonetheless, one study had a moderate take a chance of bias [27] and other evaluated studies had a low risk of bias co-ordinate to selective result reporting. Details of the quality of bias assessment are shown in Tabular array Ii.

Table II

Quality of bias cess of the included studies according to the Cochrane guidelines

Studies	Random sequence generation	Allocation concealment	Selective reporting	Blinding of participants and personnel	Blinding of upshot cess	Incomplete effect information	Other bias
Chacko, 2011 [28]	Fifty	L	L	50	50	L	L
Simental-Mendía, 2012 [45]	Fifty	L	L	50	L	Fifty	L
Kazaks, 2010 [32]	Fifty	L	Fifty	H	50	50	50
Mortazavi, 2012 [46]	L	Fifty	L	L	L	L	L
Nielsen, 2010 [29]	L	U	Fifty	Fifty	L	L	50
Resatoglu, 2004 [27]	L	H	H	L	50	Fifty	50
Rodriguez-Hernandez, 2010 [30]	Fifty	L	L	L	Fifty	50	Fifty
Simental-Mendıa, 2014 [31]	50	H	L	L	L	L	L

Characteristics of the included studies

The characteristics of included studies are summarized in Table Iii. These studies were published between 2004 and 2014 from four countries: the United states (5 studies), Mexico (iv studies), Iran and Turkey. The number of participants included in the studies ranged from xiv [28] to 100 [29]. Participants in one study were only female [thirty], while the proportion of females in five studies ranged from 29% [eight] to 78% [29]. In 2 studies the pct of women was unknown [27, 31]. The mean historic period of participants ranged from 18 [31] to 85 years [29]. The range of duration of the supplementation intervention across studies was from 8 h [27] to 6.5 months [32]. The consumed range of Mg dose in these studies was from 320 [29] to 1500 [27] mg/day. The baseline level of the CRP varied between the studies from 0.42 mg/dl [half-dozen] every bit minimum to nine.4 mg/dl [7] equally maximum, Tabular array III.

Table 3

General characteristics of the studies included

First writer, reference	Country	Study design	Inclusion criteria	Treatment duration	Sample size	Age [years]	Female (northward, %)	Mg dose	Baseline level of CRP [mg/dl]
Chacko, 2011 [28]	U.s.	Randomized, double-bullheaded, controlled, crossover trial	Being anile 30–70 years and having a trunk mass alphabetize (BMI; in kg/m2) ≥ 25, full general good health, mobility, and no dietary restrictions or allergies	4 weeks	14	thirty–70	29%	500 mg/day	1.30
Simental-Mendía, 2012 [45]	Mexico	Clinical randomized double-blind placebo-controlled trial	Impaired fasting glucose (IFG) (fasting plasma glucose levels ≥ five.5 mmol/l < 6.nine mmol/fifty), or IGT (plasma glucose levels two-h mail service-load ≥ 7.7 mmol/l < 11 mmol/50)	3 months	22	20–65	63.6%	382 mg/24-hour interval	four.ane
Kazaks, 2010 [32]	The states	Randomized placebo-controlled, double-bullheaded, parallel group intervention	–	6.5 months	52	21–55	66.6%	340 mg/twenty-four hours	3.5
Mortazavi, 2012 [46]	Islamic republic of iran	Double-blind, randomized, placebo-controlled trial	–	half dozen months	54	56.93 ±12.19	48.3%	440 mg/day	1.85
Nielsen, 2010 [29]	United states	Double-blind, placebo-controlled, supplementation trial	–	7 weeks	100	51–85	78%	320 mg/day	2.92
Resatoglu, 2004 [27]	Turkey	Randomized command trial	–	8 hours	20	–	–	1500 mg	0.42
Rodriguez-Hernandez, 2010 [30]	Mexico	Not-randomized not-placebo controlled clinical trial	–	4 months	30	30–65	100%	450 mg/day	ix.four
Simental-Mendıa, 2014 [31]	Mexico	A clinical randomized double-blind placebo-controlled trial	New diagnosis of prediabetes (glucose 5.six < seven.0 mmol/l and/or post-load glucose ≥ 7.seven < 11.1 mmol/l) and hypomagnesemia (serum magnesium levels < 0.74 mmol/50)	three months	57	eighteen–65	–	382 mg/twenty-four hours	0.72

Pooled estimate of the effect of Mg supplementation on CRP

The pooled estimate (weighted hateful difference) of the effect of Mg supplementation on CRP levels was –i.33 mg/dl (95% CI: –2.63 to –0.02, p < 0.001, heterogeneity p < 0.123; I ⁱⁱ = 29.ane%) beyond all studies (Figure 2). We divided our studies according to the baseline level of the CRP; it was constitute that subjects with baseline CRP higher than or equal to ii (CRP ≥ 2 mg/dl (four studies)) have more significant reduction in the CRP level (–2.95 mg/dl; 95% CI: –3.35 to –2.25, p < 0.001, heterogeneity p = 0.952; I ² = 1.1%) compared with the subjects with CRP < two mg/l (4 studies) (–0.23; 95% CI: –0.195 to –0.326, p < 0.001, heterogeneity p = 0.923; I ^two = 1.iii%).

Forest plot displaying weighted mean difference and 95% confidence intervals for the impact of magnesium supplementation on CRP levels

Pooled judge of the effect of Mg supplementation on IL-half dozen

The pooled estimate (weighted hateful difference) of the effect of Mg supplementation on IL-6 levels was –0.16 pg/dl (95% CI: –3.52 to three.26, p = 0.236, heterogeneity p = 0.802; I ² = 2.iii%) across all studies.

Pooled gauge of the effect of Mg supplementation on TNF-α

The pooled estimate (weighted mean difference) of the effect of Mg supplementation on TNF-α levels was 1.97 pg/dl (95% CI: 1.12 to 2.82, p = 0.043, heterogeneity p = 0.869; I ² = two.ane%) across all studies.

Pooled guess of the effect of Mg supplementation on fasting claret glucose (FBG)

The pooled estimate (weighted mean difference) of the outcome of Mg supplementation on FBG levels was –0.61 mg/dl (95% CI: –ii.72 to 1.48, p = 0.182, heterogeneity p = 0.742; I ² = 6.one%) across all studies.

Pooled estimate of the outcome of Mg supplementation on systolic blood pressure (SBP)

The pooled gauge (weighted mean difference) of the effect of Mg supplementation on SBP levels was –0.93 mm Hg (95% CI: –iii.03 to 1.xx, p = 0.293, heterogeneity p = 0.526; I ² = three.six%) beyond all studies.

Pooled estimate of the effect of Mg supplementation on diastolic claret pressure (DBP)

The pooled guess (weighted mean difference) of the outcome of Mg supplementation on DBP levels was –0.xxx mm Hg (95% CI: –ii.80 to two.19, p = 0.639, heterogeneity p = 0.489; I ⁱⁱ = iii.8%) beyond all studies.

Pooled estimate of the effect of Mg supplementation on torso mass alphabetize (BMI)

The pooled estimate (weighted hateful deviation) of the consequence of Mg supplementation on BMI levels was 0.27 kg/chiliad² (95% CI: –0.59 to 1.15, p = 0.542, heterogeneity p = 0.906; I ² = 2.0%) across all studies.

Sensitivity assay

In leave-one-out sensitivity analyses, the pooled effect estimates remained similar across all studies for CRP (–1.33 mg/50; 95% CI: –2.63 to –0.02). This stability confirms that the significant difference betwixt the studied groups is the overall effect of all included studies.

Meta regression

Random-effects meta-regression was performed to evaluate the impact of potential moderators on the estimated effect size. Changes in plasma CRP levels were independent of the dosage of Mg (slope: –0.004; 95% CI: –0.03, 0.02; p = 0.720, Figure 3) and duration (slope: –0.06; 95% CI: –0.37, 0.24; p = 0.681, Figure 4) of supplementation.

Meta-regression plots of the association between hateful changes in CRP levels with doses of magnesium supplementation. Circles represent each study, eye line is the regression line, 2 lines around the center line represent the 95% conviction interval

Meta-regression plots of the clan between hateful changes in CRP levels with duration of magnesium supplementation. Circles stand for each study, center line is the regression line, two lines around the middle line represent the 95% confidence interval

Publication bias

Visual inspection of funnel plot asymmetry suggested potential publication bias for the comparison of plasma CRP levels between Mg supplemented groups and placebo groups (Figure 5); however, the presence of publication bias was non suggested by Egger'south linear regression (intercept = –five.69, standard fault = half dozen.10; 95% CI: –xx.64, nine.25, t = 0.93, df = 6.00, two-tailed p = 0.387) and Begg's rank correlation test (Kendall's Tau with continuity correction = –0.14, z = 0.49, two tailed p = 0.620) was non indicative for publication bias. Afterwards adjustment of the consequence size for potential publication bias using the 'trim and fill' correction, no potentially missing studies were imputed in the funnel plot (WMD –one.33 mg/l; 95% CI: –2.63 to –0.02, Figure 6). The 'fail-safe N' test showed that 146 studies would be needed to bring the WMD down to a not-significant (p > 0.05) value.

Funnel plots detailing publication bias in the studies selected for analysis. Open diamond represents observed upshot size; open circles correspond observed published studies

Trim and fill method was used to impute for potentially missing studies. No potentially missing study was imputed in funnel plot, open circles represent observed published studies; open up diamond represents observed result size; closed diamond represents imputed result size

Give-and-take

The nowadays meta-analysis of 8 RCTs published in the last 10 years (2004–2014) showed that Mg supplementation significantly decreased the level of serum hs-CRP by –ane.33 mg/l (–2.63 to –0.02). Evidence from this meta-assay indicates that dietary Mg intake is inversely associated with serum CRP levels. Our results showed that Mg supplementation did not accept a significant effect on IL-six (–0.16 pg/dl). These results back up the hypothesis that dietary Mg plays a benign role in the regulation of inflammatory markers. In this regard, King [33] recently surveyed information of several cross-sectional and epidemiological studies highlighting that Mg possibly plays an of import function in potentiating inflammatory processes. The changed association constitute in our meta-analysis is supported past evidence from randomized controlled studies that take reported on patients with blazon 2 diabetes, admitted to an intensive care unit, and amongst patients with heart failure, oral Mg supplementation decreased CRP levels [34, 35]. However, previous data published regarding the efficacy of oral Mg supplementation for reducing CRP levels are deficient [35]. Regarding effects of Mg intake on IL-six level, some studies take found inconsistent results, while others have shown an changed association [36], merely others did not demonstrate any correlation [37]. IL-6, which acts as both a pro-inflammatory and an anti-inflammatory cytokine, secreted by T cells and macrophages, is the major mediator of fever and the acute stage response [38]. As raised IL-half-dozen concentrations may be an early indicator of acute inflammation, increased serum IL-vi concentrations may be observed because of undetected underlying acute infection nowadays simultaneously with Mg usage by take chances [35]. Although the Mg-related chief events that produce the acute-phase response are not known with certainty, this circuitous process suggests that Mg deficiency might be amongst the initial elements that trigger the inflammatory response [39]. Evidence derived from an animal model have shown that acute Mg deficiency leads to an inflammatory response [40]. Moreover, human studies show that low serum Mg levels are strongly associated with raised CRP concentration [3, 41]. A potential mechanism for the association between Mg deficiency and inflammation is related to calcium. If dietary Mg intake is inadequate, information technology may deplete extracellular Mg ions and consequently cause stimulation of macrophages as well as an influx of calcium ions into cells such every bit adipocytes, neuronal and peritoneal cells. N-methyl-D-aspartate receptors (NMDARs) are glutamate-gated cation channels with high calcium permeability. Augmented calcium levels in the cells enhanced Mg necessary to block the influx of calcium ions, which further led to increased stimulation of these receptors. Stimulation of these NMDAR results in the opening of channels nonselective to cations, consequently increasing calcium ions in neuronal cells [42]. This change causes the release of neurotransmitters (due east.thou., substance P) and inflammatory cytokines. IL-half dozen is released into the bloodstream and acts every bit a signaling molecule to enhance the release of CRP from the liver every bit a step in the acute phase response, which promotes prolongation of the inflammatory response in the body [i].

There are some potential limitations in our analysis that need to be addressed. Firstly, the majority of the included studies had relatively minor sample sizes, potentially leading to unstable estimates of treatment effects, because smaller trials might be methodologically less robust and are decumbent to report larger upshot sizes [43–46]. Moreover, due to the fact that the population at the baseline was very heterogeneous, we expect that it could also accept the output of the meta-analysis; nonetheless, to decrease the take a chance of bias based on the baseline level of the variables we performed a meta-regression and explored the impact of the dose and duration of the supplementation. Lastly, the number of available studies concerning the described topic was rather small, peculiarly in the case of TNF-α, which may pb to misinterpretation of the results.

In conclusion, this meta-assay suggests that Mg supplementation significantly decreased serum CRP, peculiarly with the baseline values ≥ 2 mg/dl. To provide more conclusive results and clarify the mechanistic pathways, RCTs with a larger sample size and a long-term follow-up period are warranted.

Acknowledgments

Trial registration: Systematic review registration: CRD42016039482.

MM was supported by a TWAS studentship of the Chinese Academy of Sciences, during the preparation of this manuscript.

Disharmonize of interest

The authors declare no conflict of interest.

References

1. Liu Due south, Chacko SA. Dietary Mg intake and biomarkers of inflammation and endothelial dysfunction. Magnesium in Human being Health and Affliction. Springer. 2013:35–fifty. [Google Scholar]

2. Vocal Y, Sesso HD, Manson JE, Cook NR, Buring JE, Liu S. Dietary magnesium intake and adventure of incident hypertension amongst middle-anile and older US women in a 10-year follow-upward written report. Am J Cardiol. 2006;98:1616–21. [PubMed] [Google Scholar]

3. Vocal Y, Ridker PM, Manson JE, Cook NR, Buring JE, Liu Due south. Magnesium intake, C-reactive protein, and the prevalence of metabolic syndrome in heart-aged and older US women. Diabetes Care. 2005;28:1438–44. [PubMed] [Google Scholar]

4. He K, Liu M, Daviglus ML, et al. Magnesium intake and incidence of metabolic syndrome among young adults. Circulation. 2006;113:1675–82. [PubMed] [Google Scholar]

v. Paolisso 1000, Barbagallo M. Hypertension, diabetes mellitus, and insulin resistance: the role of intracellular magnesium. Am J Hypertens. 1997;x:346–55. [PubMed] [Google Scholar]

6. Stevanović South, Nikolić 1000, Stanković A. Dietary magnesium intake and coronary centre disease risk: a study from Serbia. Medicinski Glasnik. 2011;8:203–8. [PubMed] [Google Scholar]

7. Guerrera MP, Volpe SL, Mao JJ. Therapeutic uses of magnesium. Am Fam Physician. 2009;80:157–62. [PubMed] [Google Scholar]

eight. Guerrero-Romero F, Rascón-Pacheco R, Rodríguez-Morán M, La Peña D, Escobedo J, Wacher N. Hypomagnesaemia and chance for metabolic glucose disorders: a 10-twelvemonth follow-upwards study. Eur J Clin Investig. 2008;38:389–96. [PubMed] [Google Scholar]

9. Kramer JH, Mak Information technology, Tejero-Taldo MI, et al. Neurogenic inflammation and cardiac dysfunction due to hypomagnesemia. Am J Med Sci. 2009;338:22–vii. [PMC free article] [PubMed] [Google Scholar]

10. Mazidi 1000, Kengne AP, Mikhailidis DP, Cicero AF, Banach Yard. Effects of selected dietary constituents on high-sensitivity C-reactive protein levels in U.S. adults. Ann Med. 2018;50:i–6. [PubMed] [Google Scholar]

11. King DE, Mainous AG, 3, Geesey ME, Woolson RF. Dietary magnesium and C-reactive protein levels. J Am Coll Nutr. 2005;24:166–71. [PubMed] [Google Scholar]

12. de Oliveira Otto MC, Alonso A, Lee DH, et al. Dietary micronutrient intakes are associated with markers of inflammation but not with markers of subclinical atherosclerosis. J Nutr. 2011;141:1508–xv. [PMC free commodity] [PubMed] [Google Scholar]

xiii. Moher D, Liberati A, Tetzlaff J, Altman DG. Preferred reporting items for systematic reviews and meta-analyses: the PRISMA statement. Ann Intern Med. 2009;151:264–9. [PubMed] [Google Scholar]

xiv. Phan Yard, Tian DH, Cao C, Black D, Yan TD. Systematic review and meta-analysis: techniques and a guide for the academic surgeon. Ann Cardiothorac Surg. 2015;four:112–22. [PMC free article] [PubMed] [Google Scholar]

xv. Higgins JPT GSe. Cochrane Handbook for Systematic Reviews of Interventions. Version v.0.ii. London: The Cochrane Collaboration; 2009. [Google Scholar]

16. Higgins J, Greenish Southward. Cochrane handbook for systematic reviews, version 5.0. 2 The Cochrane Collaboration. 2009. OpenURL. [Google Scholar]

17. Mazidi M, Rezaie P, Gao HK, Kengne AP. Effect of sodium-glucose cotransport-2 inhibitors on blood pressure in people with type 2 diabetes mellitus: a systematic review and meta-analysis of 43 randomized control trials with 22 528 patients. J Am Center Assoc. 2017;vi:e004007. [PMC gratis article] [PubMed] [Google Scholar]

18. Hozo SP, Djulbegovic B, Hozo I. Estimating the mean and variance from the median, range, and the size of a sample. BMC Med Res Methodol. 2005;5:13. [PMC costless article] [PubMed] [Google Scholar]

nineteen. Mazidi K, Karimi E, Rezaie P, Ferns GA. Furnishings of conjugated linoleic acid supplementation on serum C-reactive proteIn: a systematic review and meta-assay of randomized controlled trials. Cardiovasc Ther. 2017;35:e12275. [PubMed] [Google Scholar]

20. Mazidi 1000, Rezaie P, Karimi East, Kengne AP. The effects of bile acid sequestrants on lipid profile and blood glucose concentrations: a systematic review and meta-analysis of randomized controlled trials. Int J Cardiol. 2017;227:850–7. [PubMed] [Google Scholar]

21. Mazidi M, Gao HK, Rezaie P, Ferns GA. The event of ginger supplementation on serum C-reactive poly peptide, lipid profile and glycaemia: a systematic review and meta-assay. Food Nutr Res. 2016;60:32613. [PMC free commodity] [PubMed] [Google Scholar]

22. Mazidi M, Karimi E, Rezaie P, Ferns GA. Treatment with GLP1 receptor agonists reduce serum CRP concentrations in patients with type 2 diabetes mellitus: a systematic review and meta-analysis of randomized controlled trials. J Diabetes Complications. 2017;31:1237–42. [PubMed] [Google Scholar]

23. Mazidi M, Rezaie P, Vatanparast H, Kengne AP. Effect of statins on serum vitamin D concentrations: a systematic review and meta-analysis. Eur J Clin Invest. 2017;47:93–101. [PubMed] [Google Scholar]

24. Mazidi Yard, Rezaie P, Ferns GA, Gao HK. Affect of dissimilar types of tree nut, peanut, and soy nut consumption on serum C-reactive protein (CRP): a systematic review and meta-analysis of randomized controlled clinical trials. Medicine (Baltimore) 2016;95:e5165. [PMC gratis article] [PubMed] [Google Scholar]

25. Duval S, Tweedie R. Trim and fill: a simple funnel-plot-based method of testing and adjusting for publication bias in meta-analysis. Biometrics. 2000;56:455–63. [PubMed] [Google Scholar]

26. Borenstein Thousand, Hedges L, Higgins J, Rothstein H. Comprehensive Metaanalysis (Vers. 2) Englewood Cliffs, NJ: Biostat. Inc.; 2005. [Google Scholar]

27. Resatoglu AG, Demirturk O, Yener N, Yener A. Magnesium decreases cardiac injury in patients undergoing coronary artery bypass surgery. Ann Saudi Med. 2004;24:259–61. [PMC free article] [PubMed] [Google Scholar]

28. Chacko SA, Sul J, Song Y, et al. Magnesium supplementation, metabolic and inflammatory markers, and global genomic and proteomic profiling: a randomized, double-blind, controlled, crossover trial in overweight individuals. Am J Clin Nutr. 2011;93:463–73. [PMC free article] [PubMed] [Google Scholar]

29. Nielsen FH, Johnson LK, Zeng H. Magnesium supplementation improves indicators of low magnesium status and inflammatory stress in adults older than 51 years with poor quality sleep. Magnes Res. 2010;23:158–68. [PubMed] [Google Scholar]

xxx. Rodriguez-Hernandez H, Cervantes-Huerta M, Rodriguez-Moran Chiliad, Guerrero-Romero F. Oral magnesium supplementation decreases alanine aminotransferase levels in obese women. Magnes Res. 2010;23:90–half dozen. [PubMed] [Google Scholar]

31. Simental-Mendía LE, Rodríguez-Morán 1000, Guerrero-Romero F. Oral magnesium supplementation decreases C-reactive protein levels in subjects with prediabetes and hypomagnesemia: a clinical randomized double-bullheaded placebo-controlled trial. Arch Med Res. 2014;45:325–xxx. [PubMed] [Google Scholar]

32. Kazaks AG, Uriu-Adams JY, Albertson TE, Shenoy SF, Stern JS. Effect of oral magnesium supplementation on measures of airway resistance and subjective assessment of asthma control and quality of life in men and women with balmy to moderate asthma: a randomized placebo controlled trial. J Asthma. 2010;47:83–92. [PubMed] [Google Scholar]

33. Male monarch DE. Inflammation and tiptop of C-reactive protein: does magnesium play a cardinal part? Magnes Res. 2009;22:57–9. [PubMed] [Google Scholar]

34. Curiel-García JA, Rodríguez-Morán Grand, Guerrero-Romero F. Hypomagnesemia and mortality in patients with type 2 diabetes. Magnes Res. 2008;21:163–6. [PubMed] [Google Scholar]

35. Almoznino-Sarafian D, Berman S, Mor A, et al. Magnesium and C-reactive protein in heart failure: an anti-inflammatory consequence of magnesium administration? Eur J Nutr. 2007;46:230–vii. [PubMed] [Google Scholar]

36. Song Y, Li TY, van Dam RM, Manson JE, Hu FB. Magnesium intake and plasma concentrations of markers of systemic inflammation and endothelial dysfunction in women. Am J Clin Nutr. 2007;85:1068–74. [PubMed] [Google Scholar]

37. Chacko SA, Song Y, Nathan Fifty, et al. Relations of dietary magnesium intake to biomarkers of inflammation and endothelial dysfunction in an ethnically diverse accomplice of postmenopausal women. Diabetes Care. 2010;33:304–10. [PMC gratis article] [PubMed] [Google Scholar]

38. Epstein FH, Gabay C, Kushner I. Astute-phase proteins and other systemic responses to inflammation. N Engl J Med. 1999;340:448–54. [PubMed] [Google Scholar]

39. Malpuech-Brugère C, Rock E, Astier C, Nowacki W, Mazur A, Rayssiguier Y. Exacerbated immune stress response during experimental magnesium deficiency results from abnormal cell calcium homeostasis. Life Sci. 1998;63:1815–22. [PubMed] [Google Scholar]

twoscore. Malpuech-Brugère C, Nowacki W, Daveau M, et al. Inflammatory response following acute magnesium deficiency in the rat. Biochim Biophys Acta Mol Ground Dis. 2000;1501:91–8. [PubMed] [Google Scholar]

41. Guerrero-Romero F, Rodríguez-Morán M. Hypomagnesemia, oxidative stress, inflammation, and metabolic syndrome. Diabetes Metabol Res Rev. 2006;22:471–6. [PubMed] [Google Scholar]

42. Nielsen FH. Magnesium, inflammation, and obesity in chronic disease. Nutr Rev. 2010;68:333–40. [PubMed] [Google Scholar]

43. Nuesch Due east, Trelle Due south, Reichenbach S, et al. Modest study effects in meta-analyses of osteoarthritis trials: meta-epidemiological study. BMJ. 2010;341:c3515. [PMC free article] [PubMed] [Google Scholar]

44. Sterne JA, Gavaghan D, Egger M. Publication and related bias in meta-analysis: power of statistical tests and prevalence in the literature. J Clin Epidemiol. 2000;53:1119–29. [PubMed] [Google Scholar]

45. Simental-Mendía LE, Rodríguez-Morán K, Reyes-Romero MA, Guerrero-Romero F. No positive effect of oral magnesium supplementation in the decreases of inflammation in subjects with prediabetes: a pilot study. Magnes Res. 2012;25:140–six. [PubMed] [Google Scholar]

46. Mortazavi 1000, Moeinzadeh F, Saadatnia M, Shahidi South, McGee JC, Minagar A. Upshot of magnesium supplementation on carotid intima-media thickness and flow-mediated dilatation among hemodialysis patients: a double-bullheaded, randomized, placebo-controlled trial. Eur Neurol. 2013;69:309–sixteen. [PubMed] [Google Scholar]

---

Manufactures from Archives of Medical Scientific discipline : AMS are provided here courtesy of Termedia Publishing

---

mosleyqueleandon.blogspot.com

Source: https://www.ncbi.nlm.nih.gov/pmc/articles/PMC6040119/