HAND-FOOT-AND-MOUTH DISEASE EPIDEMIOLOGICAL STATUS AND RELATIONSHIP WITH METEOROLOGICAL VARIABLES IN GUANGZHOU, SOUTHERN CHINA, 2008-2012

Hand-foot-and-mouth disease (HFMD) is becoming one of the extremely common airborne and contact transmission diseases in Guangzhou, southern China, leading public health authorities to be concerned about its increased incidence. In this study, it was used an ecological study plus the negative binomial regression to identify the epidemic status of HFMD and its relationship with meteorological variables. During 2008-2012, a total of 173,524 HFMD confirmed cases were reported, 12 cases of death, yielding a fatality rate of 0.69 per 10,000. The annual incidence rates from 2008 to 2012 were 60.56, 132.44, 311.40, 402.76, and 468.59 (per 100,000), respectively, showing a rapid increasing trend. Each 1 °C rise in temperature corresponded to an increase of 9.47% (95% CI 9.36% to 9.58%) in the weekly number of HFMD cases, while a one hPa rise in atmospheric pressure corresponded to a decrease in the number of cases by 7.53% (95% CI -7.60% to -7.45%). Similarly, each one percent rise in relative humidity corresponded to an increase of 1.48% or 3.3%, and a one meter per hour rise in wind speed corresponded to an increase of 2.18% or 4.57%, in the weekly number of HFMD cases, depending on the variables considered in the model. These findings revealed that epidemic status of HFMD in Guangzhou is characterized by high morbidity but low fatality. Weather factors had a significant influence on the incidence of HFMD.

FREQUENCY OF Candida SPECIES IN A TERTIARY CARE HOSPITAL IN TRIANGULO MINEIRO, MINAS GERAIS STATE, BRAZIL

Infections by Candida species are a high-impact problem in public health due to their wide incidence in hospitalized patients. The goal of this study was to evaluate frequency, susceptibility to antifungals, and genetic polymorphism of Candida species isolated from clinical specimens of hospitalized patients. The Candida isolates included in this study were obtained from blood cultures, abdominal fluids, and central venous catheters (CVC) of hospitalized patients at the Clinical Hospital of the Federal University of Uberlândia during the period of July 2010 - June 2011. Susceptibility tests were conducted by the broth microdilution method. The RAPD-PCR tests used employed initiator oligonucleotides OPA09, OPB11, and OPE06. Of the 63 Candida isolates, 18 (28.5%) were C. albicans, 20 (31.7%) were C. parapsilosis complex species, 14 (22.2%) C. tropicalis, four (6.4%) C. glabrata, four (6.4%) C. krusei, two (3.3%) C. kefyr, and one (1.6%) C. lusitaniae. In vitro resistance to amphotericin B was observed in 12.7% of isolates. In vitroresistance to azoles was not detected, except for C. krusei. The two primers, OPA09 and OPB11, were able to distinguish different species. Isolates of C. albicans and C. parapsilosis complex species presented six and five clusters, respectively, with the OPA09 marker by RAPD-PCR, showing the genetic variability of the isolates of those species. It was concluded that members of the C. parapsilosis complex were the most frequent species found, and most isolates were susceptible to the antifungals amphotericin B, flucozanole, and itraconazole. High genetic polymorphisms were observed for isolates of C. albicans and C. parapsilosis complex species, mainly with the OPA09 marker.

PERFORMANCE OF CONVENTIONAL PCRs BASED ON PRIMERS DIRECTED TO NUCLEAR AND MITOCHONDRIAL GENES FOR THE DETECTION AND IDENTIFICATION OF Leishmania spp.

In visceral leishmaniasis, the detection of the agent is of paramount importance to identify reservoirs of infection. Here, we evaluated the diagnostic attributes of PCRs based on primers directed to cytochrome-B (cytB), cytochrome-oxidase-subunit II (coxII), cytochrome-C (cytC), and the minicircle-kDNA. Although PCRs directed to cytB, coxII, cytC were able to detect different species of Leishmania, and the nucleotide sequence of their amplicons allowed the unequivocal differentiation of species, the analytical and diagnostic sensitivity of these PCRs were much lower than the analytical and diagnostic sensitivity of the kDNA-PCR. Among the 73 seropositive animals, the asymptomatic dogs had spleen and bone marrow samples collected and tested; only two animals were positive by PCRs based on cytB, coxII, and cytC, whereas 18 were positive by the kDNA-PCR. Considering the kDNA-PCR results, six dogs had positive spleen and bone marrow samples, eight dogs had positive bone marrow results but negative results in spleen samples and, in four dogs, the reverse situation occurred. We concluded that PCRs based on cytB, coxII, and cytC can be useful tools to identify Leishmania species when used in combination with automated sequencing. The discordance between the results of the kDNA-PCR in bone marrow and spleen samples may indicate that conventional PCR lacks sensitivity for the detection of infected dogs. Thus, primers based on the kDNA should be preferred for the screening of infected dogs.

HIV genotyping among female sex workers in the State of Santa Catarina

The objective of this study was to investigate the frequency of HIV infection among female sex workers in the port area of Imbituba (State of Santa Catarina), and to identify the viral subtype and its susceptibility to antiretroviral medications. Ninety women were interviewed between December 2003 and February 2004. Six (6.7%) were HIV-positive. Genotyping for HIV, performed on four samples, detected subtype C in three of them, which is predominant in Africa and Asia, and subtype B in one of them, which is prevalent in Brazil, USA and Europe. The results suggest that the Port of Imbituba may be one of the gateways for HIV-1 subtype C to enter Brazil, and for its dissemination to the rest of the country and the Mercosul area, along the highway BR-101. This points towards the need for preventive work to reduce the introduction and dissemination of HIV subtype C in Brazil.

Larvicidal activity of oil-resin fractions from the Brazilian medicinal plant Copaifera reticulata Ducke (Leguminosae-Caesalpinoideae) against Aedes aegypti (Diptera, Culicidae)

Oil-resin fractions from Copaifera reticulata Ducke (Leguminosae-Caesalpinoideae) were evaluated for larvicidal activity on third larval instars of Aedes aegypti, in searching for alternative control methods for this mosquito. The bioactive fractions were chemically monitored by thin-layer chromatography, ¹H and 13C nuclear magnetic resonance and mass spectrometry. Bioassays were performed using five repetitions, at a temperature of 28 ± 1°C, relative humidity of 80 ± 5% and light and dark cycles of 12h. Mortality was indicated by darkening of the cephalic capsule after 24h of exposure of the larvae to the solutions. The most active fractions were CRM1-4 (sesquiterpenes) and CRM5-7 (labdane diterpenes), which showed LC50 values of 0.2 and 0.8ppm, respectively.

Temperature and atmospheric pressure may be considered as predictors for the occurrence of bacillary dysentery in Guangzhou, Southern China

Introduction The control of bacillary dysentery (BD) remains a big challenge for China. Methods Negative binomial multivariable regression was used to study relationships between meteorological variables and the occurrence of BD during the period of 2006-2012. Results Each 1°C rise of temperature corresponded to an increase of 3.60% (95%CI, 3.03% to 4.18%) in the monthly number of BD cases, whereas a 1 hPa rise in atmospheric pressure corresponded to a decrease in the number of BD cases by 2.85% (95%CI = 3.34% to 2.37% decrease). Conclusions Temperature and atmospheric pressure may be considered as predictors for the occurrence of BD in Guangzhou.

Efeito da aplicação de calcário e micronutrientes em latossolo amarelo da Amazônia Central

O efeito de diferentes níveis de colagem na presença de micronutrientes, foram testados na cultuha da soja, em Latossolo Amarelo da Amazônia Central, sob três áreas com diferentes maneiras de preparo do solo: desmatamento seguido da queima de floresta primária; desmatamento seguido da retirada da floresta primária com aso de maquinas e desmatamento seguido da queima de vegetação de capoeira. As produções obtidas com as doses 1, 3 c 5 ton/ka, na ausência de micronutrientes, não diferiram estatísticamente da testemunha (0 ton/ha). No entanto, com a aplicação de 2 ton/calcário mais micronutrientes, o aumento de produção foi bastante pronunciado. 0s menores rendimentos foram obtidos no solo preparo com o uso de máquinas. A disponibilidade de nutrientes em Latossolo Amarelo, assim como o preparo de solo é discutido.

Effects of a novel fermented soy product on the serum lipids of hypercholesterolemic rabbits

OBJECTIVE: To assess the effect of a new feed soy product fermented by Enterococcus faecium and Lactobacillus jugurti on the serum lipid levels of rabbits with induced hypercholesterolemia. METHODS: Thirty-two rabbits were divided into 4 groups as follows: 1) control (C); 2) hypercholesterolemic (H); 3) hypercholesterolemic + fermented product (HPF); and 4) control + fermented product (CPF). The H and HPF groups were fed with a diet with 0.15% (p/p) cholesterol in the first 15 days. C and CPF groups received regular food preparation. The HPF and CPF groups received 10 mL daily of the fermented 30 days. Blood samples were drawn at the beginning of the study and at the 15th and 30th days. Concentrations of total cholesterol, HDL-cholesterol, and triglycerides were analyzed. RESULTS: After 15 days, the HPF group showed a total cholesterol concentration lower (18.4%) than that of the H group (p=0.05), but this difference disappeared after 30 days. No change was observed in total cholesterol levels of C and CPF groups. After 15 days, the HDL-cholesterol was higher (17.8%) in the HPF group, but the triglyceride levels remained unchanged in all groups during the same period of time. CONCLUSION: The soy fermented product caused an 18.4% reduction in total cholesterol and a 17.8% increase in the HDL-fraction. It may, therefore, be a possible coadjutor in the treatment of hypercholesterolemia.

Efeitos do betacaroteno e do tabagismo sobre a remodelação cardíaca pós-infarto do miocárdio

OBJETIVO: Analisar os efeitos do betacaroteno no processo de remodelação ventricular após o infarto agudo do miocárdio (IAM), em ratos expostos à fumaça do cigarro. MÉTODOS: Após o IAM, os animais foram divididos em quatro grupos: 1) grupo C, 24 animais que receberam dieta-padrão; 2) grupo BC, 26 animais que receberam betacaroteno; 3) grupo EFC, 26 animais que receberam dieta-padrão e foram expostos à fumaça de cigarro; e 4) grupo BC+EFC, 20 animais que receberam betacaroteno e foram expostos à fumaça de cigarro. Após seis meses, foi realizado estudo morfofuncional. Utilizou-se significância de 5%. RESULTADOS: Em relação às áreas diastólicas (AD) e sistólicas (AS), os valores do grupo BC foram maiores que os do grupo C. Considerando a AD/peso corporal (PC) e AS/PC, os valores do grupo BC+EFC foram maiores que os valores de C. Em relação à fração de variação de área, foram observadas diferenças significativas entre EFC (valores menores) e C (valores maiores) e entre BC (valores menores) e C (valores maiores). Não foram observadas diferenças entre os grupos em relação ao tamanho do infarto. O grupo EFC apresentou valores maiores da área seccional dos miócitos (ASM) que os animais-controle. Em adição, o grupo BC+EFC apresentou maiores valores de ASM que BC, EFC e C. CONCLUSÃO: Após o infarto do miocárdio, o tabagismo e o betacaroteno promoveram intensificação do processo de remodelação cardíaca; houve potencialização dos efeitos deletérios no processo de remodelação com os dois tratamentos em conjunto.

Sex-Specific Equations to Estimate Maximum Oxygen Uptake in Cycle Ergometry

AbstractBackground:Aerobic fitness, assessed by measuring VO2max in maximum cardiopulmonary exercise testing (CPX) or by estimating VO2max through the use of equations in exercise testing, is a predictor of mortality. However, the error resulting from this estimate in a given individual can be high, affecting clinical decisions.Objective:To determine the error of estimate of VO2max in cycle ergometry in a population attending clinical exercise testing laboratories, and to propose sex-specific equations to minimize that error.Methods:This study assessed 1715 adults (18 to 91 years, 68% men) undertaking maximum CPX in a lower limbs cycle ergometer (LLCE) with ramp protocol. The percentage error (E%) between measured VO2max and that estimated from the modified ACSM equation (Lang et al. MSSE, 1992) was calculated. Then, estimation equations were developed: 1) for all the population tested (C-GENERAL); and 2) separately by sex (C-MEN and C-WOMEN).Results:Measured VO2max was higher in men than in WOMEN: -29.4 ± 10.5 and 24.2 ± 9.2 mL.(kg.min)-1 (p < 0.01). The equations for estimating VO2max [in mL.(kg.min)-1] were: C-GENERAL = [final workload (W)/body weight (kg)] x 10.483 + 7; C-MEN = [final workload (W)/body weight (kg)] x 10.791 + 7; and C-WOMEN = [final workload (W)/body weight (kg)] x 9.820 + 7. The E% for MEN was: -3.4 ± 13.4% (modified ACSM); 1.2 ± 13.2% (C-GENERAL); and -0.9 ± 13.4% (C-MEN) (p < 0.01). For WOMEN: -14.7 ± 17.4% (modified ACSM); -6.3 ± 16.5% (C-GENERAL); and -1.7 ± 16.2% (C-WOMEN) (p < 0.01).Conclusion:The error of estimate of VO2max by use of sex-specific equations was reduced, but not eliminated, in exercise tests on LLCE.

A determinação dos números de indivíduos mínimos necessários na experimentação genética

The main object of the present paper consists in giving formulas and methods which enable us to determine the minimum number of repetitions or of individuals necessary to garantee some extent the success of an experiment. The theoretical basis of all processes consists essentially in the following. Knowing the frequency of the desired p and of the non desired ovents q we may calculate the frequency of all possi- ble combinations, to be expected in n repetitions, by expanding the binomium (p-+q)n. Determining which of these combinations we want to avoid we calculate their total frequency, selecting the value of the exponent n of the binomium in such a way that this total frequency is equal or smaller than the accepted limit of precision n/pª{ 1/n1 (q/p)n + 1/(n-1)| (q/p)n-1 + 1/ 2!(n-2)| (q/p)n-2 + 1/3(n-3) (q/p)n-3... < Plim - -(1b) There does not exist an absolute limit of precision since its value depends not only upon psychological factors in our judgement, but is at the same sime a function of the number of repetitions For this reasen y have proposed (1,56) two relative values, one equal to 1-5n as the lowest value of probability and the other equal to 1-10n as the highest value of improbability, leaving between them what may be called the "region of doubt However these formulas cannot be applied in our case since this number n is just the unknown quantity. Thus we have to use, instead of the more exact values of these two formulas, the conventional limits of P.lim equal to 0,05 (Precision 5%), equal to 0,01 (Precision 1%, and to 0,001 (Precision P, 1%). The binominal formula as explained above (cf. formula 1, pg. 85), however is of rather limited applicability owing to the excessive calculus necessary, and we have thus to procure approximations as substitutes. We may use, without loss of precision, the following approximations: a) The normal or Gaussean distribution when the expected frequency p has any value between 0,1 and 0,9, and when n is at least superior to ten. b) The Poisson distribution when the expected frequecy p is smaller than 0,1. Tables V to VII show for some special cases that these approximations are very satisfactory. The praticai solution of the following problems, stated in the introduction can now be given: A) What is the minimum number of repititions necessary in order to avoid that any one of a treatments, varieties etc. may be accidentally always the best, on the best and second best, or the first, second, and third best or finally one of the n beat treatments, varieties etc. Using the first term of the binomium, we have the following equation for n: n = log Riim / log (m:) = log Riim / log.m - log a --------------(5) B) What is the minimun number of individuals necessary in 01der that a ceratin type, expected with the frequency p, may appaer at least in one, two, three or a=m+1 individuals. 1) For p between 0,1 and 0,9 and using the Gaussean approximation we have: on - ó. p (1-p) n - a -1.m b= δ. 1-p /p e c = m/p } -------------------(7) n = b + b² + 4 c/ 2 n´ = 1/p n cor = n + n' ---------- (8) We have to use the correction n' when p has a value between 0,25 and 0,75. The greek letters delta represents in the present esse the unilateral limits of the Gaussean distribution for the three conventional limits of precision : 1,64; 2,33; and 3,09 respectively. h we are only interested in having at least one individual, and m becomes equal to zero, the formula reduces to : c= m/p o para a = 1 a = { b + b²}² = b² = δ2 1- p /p }-----------------(9) n = 1/p n (cor) = n + n´ 2) If p is smaller than 0,1 we may use table 1 in order to find the mean m of a Poisson distribution and determine. n = m: p C) Which is the minimun number of individuals necessary for distinguishing two frequencies p1 and p2? 1) When pl and p2 are values between 0,1 and 0,9 we have: n = { δ p1 ( 1-pi) + p2) / p2 (1 - p2) n= 1/p1-p2 }------------ (13) n (cor) We have again to use the unilateral limits of the Gaussean distribution. The correction n' should be used if at least one of the valors pl or p2 has a value between 0,25 and 0,75. A more complicated formula may be used in cases where whe want to increase the precision : n (p1 - p2) δ { p1 (1- p2 ) / n= m δ = δ p1 ( 1 - p1) + p2 ( 1 - p2) c= m / p1 - p2 n = { b2 + 4 4 c }2 }--------- (14) n = 1/ p1 - p2 2) When both pl and p2 are smaller than 0,1 we determine the quocient (pl-r-p2) and procure the corresponding number m2 of a Poisson distribution in table 2. The value n is found by the equation : n = mg /p2 ------------- (15) D) What is the minimun number necessary for distinguishing three or more frequencies, p2 p1 p3. If the frequecies pl p2 p3 are values between 0,1 e 0,9 we have to solve the individual equations and sue the higest value of n thus determined : n 1.2 = {δ p1 (1 - p1) / p1 - p2 }² = Fiim n 1.2 = { δ p1 ( 1 - p1) + p1 ( 1 - p1) }² } -- (16) Delta represents now the bilateral limits of the : Gaussean distrioution : 1,96-2,58-3,29. 2) No table was prepared for the relatively rare cases of a comparison of threes or more frequencies below 0,1 and in such cases extremely high numbers would be required. E) A process is given which serves to solve two problemr of informatory nature : a) if a special type appears in n individuals with a frequency p(obs), what may be the corresponding ideal value of p(esp), or; b) if we study samples of n in diviuals and expect a certain type with a frequency p(esp) what may be the extreme limits of p(obs) in individual farmlies ? I.) If we are dealing with values between 0,1 and 0,9 we may use table 3. To solve the first question we select the respective horizontal line for p(obs) and determine which column corresponds to our value of n and find the respective value of p(esp) by interpolating between columns. In order to solve the second problem we start with the respective column for p(esp) and find the horizontal line for the given value of n either diretly or by approximation and by interpolation. 2) For frequencies smaller than 0,1 we have to use table 4 and transform the fractions p(esp) and p(obs) in numbers of Poisson series by multiplication with n. Tn order to solve the first broblem, we verify in which line the lower Poisson limit is equal to m(obs) and transform the corresponding value of m into frequecy p(esp) by dividing through n. The observed frequency may thus be a chance deviate of any value between 0,0... and the values given by dividing the value of m in the table by n. In the second case we transform first the expectation p(esp) into a value of m and procure in the horizontal line, corresponding to m(esp) the extreme values om m which than must be transformed, by dividing through n into values of p(obs). F) Partial and progressive tests may be recomended in all cases where there is lack of material or where the loss of time is less importent than the cost of large scale experiments since in many cases the minimun number necessary to garantee the results within the limits of precision is rather large. One should not forget that the minimun number really represents at the same time a maximun number, necessary only if one takes into consideration essentially the disfavorable variations, but smaller numbers may frequently already satisfactory results. For instance, by definition, we know that a frequecy of p means that we expect one individual in every total o(f1-p). If there were no chance variations, this number (1- p) will be suficient. and if there were favorable variations a smaller number still may yield one individual of the desired type. r.nus trusting to luck, one may start the experiment with numbers, smaller than the minimun calculated according to the formulas given above, and increase the total untill the desired result is obtained and this may well b ebefore the "minimum number" is reached. Some concrete examples of this partial or progressive procedure are given from our genetical experiments with maize.

Sôbre uma propriedade da equação utilizada para interpolação da Lei de Mitscherlich

The author proves that equation, Σy n ΣZx | ΣxyZx ΣxZx ΣxZ2x | = 0, Σy ΣZx Σy2x | where Z = 10-cq and q is a numerical constant, used by Pimentel Gomes and Malavolta in several articles for the interpolation of Mitscherlih's equation y = A [ 1 - 10 - c (x + b) ] by the least squares method, always has a zero of order three for Z = 1. Therefore, equation A Zm + A1Zm -1 + ........... + Am = 0 obtained from that determinant can be divided by (Z-1)³. This property provides a good test for the correctness of the computations and facilitates the solution of the equation.

Um caso especial de integrais binômias

The present paper shows that the sum of two binomial integrals, such as A ∫ x p (a + bx q)r dx + B ∫ x p (a + bx q)r dx, where A and B are real constants and p, q, r are rational numbers, can, in special cases, lead to elementary integrals, even if each by itself is not elementary. An example of the case considered is given by the integral ∫ x _____-___ 3 dx = 1/2 ∫ x-½ (x - 1)-⅓ dx - 6 √ x ³√(x - 1)4 = 1/3 ∫ x-½ (x - 1)-¾ dx On the rigth hand side of the last equality both integral are not elementary. But the use of integration by parts of one of them leads to the solution: ∫ x _____-___ 3 dx = x½ (x - 1)-⅓ + C. 6 √ x ³√(x - 1)4

O problema técnico-econômico da adubação

The authors discuss from the economic point of view the use of a few functions intended to represent the yield y corresponding to a level xof the nutrient. They point out that under conditions of scarce capital what is actually most important is not to obtain the highest profit per hectare but the highest return per cruzeiro spent, so that we should maximize the function z = _R - C_ = _R_ - 1 , C C where R is the gross income and C the cost of production (fixed plus variable, both per hectare). Being C = M + rx, with r the unit price of the nutrient and Af the fixed cost of the crop, wo are led to the equation (M + rx)R' - rR = 0. With R = k + sx + tx², this gives a solution Xo = - Mt - √ M²t² - r t(Ms - Kr)- _____________________ rt on the other hand, with R = PyA [1 - 10-c(x + b)], x0 will be the root of equation (M + rx)cL 10 + r 10c(x + b) = 0 (12). Another solution, pointed out by PESEK and HEADY, is to maximize the function z = sx + tx² _________ m + rx where the numerator is the additional income due to the nutrient, and m is the fixed cost of fertilization. This leads to a solution x+ = - mt - √m²t² - mrst (13) _________________ rt However, we must have x+< _r_-_s_ I if we want to satisfy t _dy_ > r. dx This condition is satisfied only if we have m < _(s__-__r)² (14), - 4 t a restriction apparently not perceived by PESEK and HEADY. A similar reasoning using Mitscherlich's law leads to equation (mcL 10 + r) + cr(L 10)x - r 10cx = 0 (15), with a similar restriction. As an example, data of VIEGAS referring to fertilization of corn (maize) gave the equation y - 1534 + 22.99 x - 0. 1069 x², with x in kg/ha of the cereal. With the prices of Cr$ 5.00 per kilo of maize, Cr$ 26.00 per kilo of P2O3,. and M = Cr$ 5,000.00, we obtain x0 = 61 kg/ha of P(2)0(5). A similar reasoning using Mitscherlich's law leads to x0 = 53 kg/ha. Now, if we take in account only the fixed cost of fertilization m = Cr$ 600.00 per hectare, we obtain from (13) x+ = 51 kg/ha of P2O5, while (14) gives x+ - 41 kg/ha. Note that if m = Cr$ 5,000.00, we obtain by formula (13) x+ = 88 kg/ha of P2O5, a solution which is not valid, since condition (14) is not satisfied.

Factors affecting phosphate fixation in soils of the state of São Paulo

A laboratory experiment was carried out to study the effects of chemical and physical characteristics of the soil on the phosphate fixing capacity. One hundred samples collected from various localities were at first characterized chemically and their particlesize distribution determined. They were then tested as to their phosphate fixing capacities. The results obtained were statistically analysed by means of both simple linear and multiple correlation. The following conclusions could be drawn: 1. simple linear regression analysis indicated that % C, exchangeable Al+3, CEC, % clay, pH and % sand were the soil characteristics which significantly affected phosphate fixing capacity of São Paulo State soils; 2. multiple linear regression analysis indicated that % C, exchangeable Mg(+2)9 exchangeable- Al+3 and % clay were the soil characteristics which significantly affected the phosphate fixing capacity of São Paulo State soils; 3. the phosphate phonomena fixing as they occur in the soils of the São Paulo State can be best described by the following equation: Y = -2,266 - 3,484 + 3,514 + 5,559 + 1,005 %C Mg+2 Al+3 % clay exchangeable exchangeable 4. phosphate fixation in the soil is affected by the combined effects of both soil chemical and physical characteristics.