Distribution of cardiac geometric patterns on echocardiography in essential hypertension. Impact of two criteria of stratification

PURPOSE: To evaluate 2 left ventricular mass index (LVMI) normality criteria for the prevalence of left ventricular geometric patterns in a hypertensive population ( HT ) . METHODS: 544 essential hypertensive patients, were evaluated by echocardiography, and different left ventricular hypertrophy criteria were applied: 1 - classic : men - 134 g/m² and women - 110 g/m² ; 2- obtained from the 95th percentil of LVMI from a normotensive population (NT). RESULTS: The prevalence of 4 left ventricular geometric patterns, respectively for criteria 1 and 2, were: normal geometry - 47.7% and 39.3%; concentric remodelying - 25.4% and 14.3%; concentric hypertrophy - 18.4% and 27.7% and excentric hypertrophy - 8.8% and 16.7%, which confered abnormal geometry to 52.6% and 60.7% of hypertensive. The comparative analysis between NT and normal geometry hypertensive group according to criteria 1, detected significative stuctural differences,"( *p < 0.05):LVMI- 78.4 ± 1.50 vs 85.9 ±0.95 g/m² *; posterior wall thickness -8.5 ± 0.1 vs 8.9 ± 0.05 mm*; left atrium - 33.3 ± 0.41 vs 34.7 ± 0.30 mm *. With criteria 2, significative structural differences between the 2 groups were not observed. CONCLUSION: The use of a reference population based criteria, increased the abnormal left ventricular geometry prevalence in hypertensive patients and seemed more appropriate for left ventricular hypertrophy detection and risk stratification.

Distribution of risk factors in parents and siblings of patients with early coronary artery disease

OBJECTIVE: Early coronary artery disease (CAD) is associated with risk factors (RF). Offspring of parents with a RF have a greater prevalence of them. However, the distribution of RF in parents and siblings of patients with early CAD is unknown. METHODS: The study comprised the parents and siblings of 42 patients with early CAD (< 45 years), 29 males. Their mean age was 39.5±3.7 years. The following major RF were analyzed: smoking (> 5 cigarretes/day), hypercholesterolemia (total cholesterol > 200 mg/dL), hypertension (diastolic blood pressure > 90 mmHg), and diabetes (glycemia > 126 mg/dL). RESULTS: Of a total of 102 RF, 4, 3, 2, and 1 were observed in, respectively, 5, 15, 15, and 7 patients with early CAD, the most prevalent being smoking (86%) and hypercholesterolemia (83%). Diabetes was observed in 15 (36%) and hypertension in 16 (38%) patients. Smoking was more prevalent in the fathers (76%) and hypercholesterolemia in the mothers (30%). In 183 siblings, 131 RF were observed (1 patient with the disease had a mean of 4.7 siblings). The prevalences of smoking, hypertension, hypercholesterolemia, and diabetes in the siblings were, respectively, 32%, 18%, 14%, and 9%. The incidence of RF was as follows: 72 (39%) siblings had 1 RF, 25 (14%) siblings had 2 RF, and 3 (2%) siblings had 3 RF. In parents and their offspring, smoking was moderately correlated (r=0.43; P=0.02) with CAD. CONCLUSION: Smoking habit of parents is passed on to offspring, and, in association with hypercholesterolemia, it was the major cause of early CAD in offspring. High prevalence of smoking in offspring shows the potential responsibility of parents in the incidence of the disease in offspring.

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.

Estudo comparativo entre métodos de contrôle quantitativo da produção leiteira

This paper deals with the estimation of milk production by means of weekly, biweekly, bimonthly observations and also by method known as 6-5-8, where one observation is taken at the 6th week of lactation, another at 5th month and a third one at the 8th month. The data studied were obtained from 72 lactations of the Holstein Friesian breed of the "Escola Superior de Agricultura "Luiz de Queiroz" (Piracicaba), S. Paulo, Brazil), being 6 calvings on each month of year and also 12 first calvings, 12 second calvings, and so on, up to the sixth. The authors criticize the use of "maximum error" to be found in papers dealing with this subject, and also the use of mean deviation. The former is completely supersed and unadvisable and latter, although equivalent, to a certain extent, to the usual standard deviation, has only 87,6% of its efficiency, according to KENDALL (9, pp. 130-131, 10, pp. 6-7). The data obtained were compared with the actual production, obtained by daily control and the deviations observed were studied. Their means and standard deviations are given on the table IV. Inspite of BOX's recent results (11) showing that with equal numbers in all classes a certain inequality of varinces is not important, the autors separated the methods, before carrying out the analysis of variance, thus avoiding to put together methods with too different standard deviations. We compared the three first methods, to begin with (Table VI). Then we carried out the analysis with the four first methods. (Table VII). Finally we compared the two last methods. (Table VIII). These analysis of variance compare the arithmetic means of the deviations by the methods studied, and this is equivalent to compare their biases. So we conclude tht season of calving and order of calving do not effect the biases, and the methods themselves do not differ from this view point, with the exception of method 6-5-8. Another method of attack, maybe preferrable, would be to compare the estimates of the biases with their expected mean under the null hypothesis (zero) by the t-test. We have: 1) Weekley control: t = x - 0/c(x) = 8,59 - 0/ = 1,56 2) Biweekly control: t = 11,20 - 0/6,21= 1,80 3) Monthly control: t = 7,17 - 0/9,48 = 0,76 4) Bimonthly control: t = - 4,66 - 0/17,56 = -0,26 5) Method 6-5-8 t = 144,89 - 0/22,41 = 6,46*** We denote above by three asterisks, significance the 0,1% level of probability. In this way we should conclude that the weekly, biweekly, monthly and bimonthly methods of control may be assumed to be unbiased. The 6-5-8 method is proved to be positively biased, and here the bias equals 5,9% of the mean milk production. The precision of the methods studied may be judged by their standard deviations, or by intervals covering, with a certain probability (95% for example), the deviation x corresponding to an estimate obtained by cne of the methods studied. Since the difference x - x, where x is the mean of the 72 deviations obtained for each method, has a t distribution with mean zero and estimate of standard deviation. s(x - x) = √1+ 1/72 . s = 1.007. s , and the limit of t for the 5% probability, level with 71 degrees of freedom is 1.99, then the interval to be considered is given by x ± 1.99 x 1.007 s = x ± 2.00. s The intervals thus calculated are given on the table IX.

Doses and methods of distribution of potassium chloride on soybean crop (Glycine max (L.) Merrill) in a red latosol, affecting soil salinity, grain production and chemical composition of leaves

The present work deal t wi th an experiment under field conditions and a laboratory test of soil incubation the objectives were as follows: a. to study effects on soybean grain product ion and leaf composition of increasing doses of potassium chloride applied into the soil through two methods of distribution; b. to observe chemical modifications in the soils incubated with increasing doses of potassium chloride; and, c. to correlate field effects with chemical alterations observed in the incubation test, The field experiment was carried out in a Red Latosol (Haplustox) with soybean cultivar UFV - 1. Potassium chloride was distributed through two methods: banded (5 cm below and 5 cm aside of the seed line) and broadcasted and plowed-down. Doses used were: 0; 50; 100 and 200 kg/ha of K2O. Foliar samples were taken at flowering stage. Incubation test were made in plastic bags with 2 kg of air dried fine soil, taken from the arable layer of the field experiment, with the following doses of KC1 p,a. : 0; 50; 100; 200; 400; 800; 1,600; 3.200; 6,400 and 12,800 kg/ha of K(2)0. In the conditions observed during the present work, results allowed the following conclusions: A response by soybean grain production for doses of potassium chloride, applied in both ways, banded or broadcasted, was not observed. Leaf analysis did not show treatment influence over the leaf contents for N, P, K, Ca, Mg, and CI, Potassium chloride salinity effects in both methods of distribution for all the tested closes were not observed.

The cassava mite complex: III - new distribution records, mainly from Colombia and Africa. References to other plants

Uma lista de novas referências e ocorrências para ácaros tetraniquídeos da mandioca é apresentada.

Distribution and morphometrics of Natalus stramineus from South America (Chiroptera, Natalidae)

Morphometric and distributional data and some observations on the biology of Natalus stramineus Gray, 1838 collected in eastern Bolivia and in northern, northeastern, central, and southeastern Brazil are presented. All new records, combined with the records of the species from Paraguay and Mato Grosso, significantly change the known distribution of N. stramineus in South America. The specimens from northeastern Brazil (Rio Grande do Norte, Ceará, Bahia) are smaller than those found in the northern (Pará), eastern (Espírito Santo, São Paulo) and central regions of the country (Distrito Federal, Goiás, Mato Grosso do Sul). Natalus stramineus specimens from the three latter regions are about the same size, but are larger than those from Santa Cruz, Bolivia. Their size is intermediate between those of central samples and northeastern Brazil samples. The type locality of this species is discussed.

Sewage impact on the composition and distribution of Polychaeta associated to intertidal mussel beds of the Mar del Plata rocky shore, Argentina

The polychaete composition and distribution within mussel beds were studied in order to assess organic pollution due to domestic sewage in a rocky shore of Mar del Plata (Argentina) during 1997. Four stations and a control site were randomly sampled around the local effluent. Quantitative data on polychaetes, as well as sediment accumulated among mussels and its organic carbon content were measured. Polychaete distribution patterns are related to the organic matter gradient, being Capitella cf. capitata, Neanthes succinea (Frey & Leuckart, 1847) and Boccardia polybranchia (Haswell, 1885) the dominant indicator species close to the effluent. At medial distances, the cirratulids Caulleriella alata (Southern, 1914) and Cirratulus cirratus (Müller, 1776) are very important in abundance. The syllids Syllis prolixa Ehlers, 1901 and S. gracilis Grube, 1840 are distributed along the study area, but dominate at the medial stations and at the control site. The orbiniid Protoariciella uncinata Hartmann-Schröder, 1962 is subdominant at the control station.

Chaetognatha of the Brazil-Malvinas (Falkland) confluence: distribution and associations

The planktonic chaetognaths from the Brazil-Malvinas (Falkland) confluence, extending between 36º 30' - 50º 5' S and 60º 33' - 41º 7' W, were studied. Ten species were found: Eukrohnia hamata (Möbius, 1875) (Eukrohniidae), Pterosagitta draco (Krohn, 1853) (Pterosagittidae), Sagitta enflata Grassi, 1881, Sagitta gazellae Ritter-Zahony, 1909, Sagitta hexaptera d´Orbigny, 1834, Sagitta lyra Krohn, 1853, Sagitta minima Grassi, 1881, Sagitta planctonis Steinhaus, 1896, Sagitta serratodentata Krohn, 1853, and Sagitta tasmanica Thomson, 1947 (Sagittidae). Sagitta gazellae was the most abundant species followed by E. hamata, S. tasmanica and S. serratodentata. The association analysis among the different species, salinity and temperature revealed two groups of species, one related to higher salinities and warmer waters (P. draco, S. hexaptera and S. serratodentata) and the other to lower salinities and colder waters (E. hamata, S. gazellae and S. tasmanica). The fact that P. draco and S. hexaptera, formerly defined as warm-water species, appeared further south than previously reported might be related to the existence of warm core eddies up to 46º S in September and October 1988.

Spatial and temporal distribution of decapod larvae in the subtropical waters of the Arvoredo archipelago, SC, Brazil

The present paper aims to describe the temporal and spatial distribution of the composition and abundance of Decapoda larvae in the shallow waters around Arvoredo Marine Biological Reserve. Stomatopod occurrence is also discussed. Plankton samples were collected at five sites around the Arvoredo Island every two months for one year from May, 2002 to April, 2003. Thirty-nine morphotypes, 11 genus and 4 species (Artemesia longinaris Bate, 1888, Hexapanopeus schmitii Rathbun, 1930, Menippe nodifrons Stimpson, 1859 and Pleoticus muelleri Bate, 1888) were identified, among them only two morphotypes of Stomatopoda larvae, and the remainder Decapoda larvae. Brachyuran zoeae were the most abundant group and they were well represented by Portunidae and Xanthidae zoeae. Lucifer sp. and Caridea zoeae were the most abundant non-brachyuran taxa. Decapod larvae were observed to occur at all sampling sites, however the spatial distribution demonstrated a general tendency to greater abundance and diversity at the southern sites of the Island. Decapoda and Stomatopoda larvae occurred throughout the year, showing that reproduction is continuous, but that larval input in planktonic community was significantly higher during autumn and spring.

Ostracodes (Crustacea) from Cananéia-Iguape estuarine/lagoon system and geographical distribution of the mixohaline assemblages in southern and southeastern Brazil

The ostracode assemblages from Cananéia-Iguape estuarine/lagoon system (southernmost State of São Paulo) are here discussed in detail for the first time. Thirty-four sites, approximately 1 km equidistant, were sampled along the system, including the Cananéia Sea, Pequeno Sea, Cubatão Sea, Ribeira de Iguape River and Itapitangui River. The ostracodes throughout this area have poor assemblages, with a total of 662 specimens of dead and living organisms. The majority of the ostracode fauna is composed of euryhaline species, as follows: Cyprideis multidentata Hartmann, 1955 (174 specimens), Minicythere heinii Ornellas, 1974 (54 specimens), Tanella gracilis Kingma, 1948 (96 specimens) and Whatleyella sanguinettiae Coimbra, Carreño & Ferron, 1994 (226 specimens). Although there are few studies on the Brazilian mixohaline ostracode faunas, including the euryhaline marginal marine taxa, the published data show that the group is best known in the south and southeast regions. Based on this review and with the new data presented in this paper, the geographical distribution of eight mixohaline key species in southern and southeastern Brazil is also discussed.

Geographic distribution and morphological variation in Mimon bennettii (Chiroptera, Phyllostomidae)

We studied discrete and quantitative data from 88 specimens of the subgenus Mimon previously identified as Mimon bennettii (Gray, 1838) and M. cozumelae Goldman, 1914 from diverse parts of their range. Our data indicate that specimens of Mimon bennetii in Brazil presented geographic variation in morphometrical characters and mosaic variation in qualitative traits. Specimens from the Cerrado biome collected in Brazilian states like Piaui, Tocantins, and Goiás have longer forearms than those distributed in the Atlantic and Amazon forested domains. Based on morphometrics, as showed by t-tests, specimens of M. bennettii from the Brazilian Cerrado resemble phenetically more with M. cozumelae than the M. bennettii from Atlantic Forest. Characters presently used to diagnosis M. cozumelae also were also recorded to M. bennettii in diverse parts of Brazil, making that validity of M. cozumelae questionable based on this kind of traits. This research also updated the geographic distribution to the M. bennettii in Brazil.

Spatial and temporal distribution of Pseudis minuta (Anura, Hylidae, Hylinae) and environmental variables related to its reproductive activity in Reserva Biológica do Lami, southern Brazil

We studied the reproductive biology of a population of Pseudis minuta Günther, 1858 from Reserva Biológica do Lami (30º 15' S; 51º 05' W), Porto Alegre, southern Brazil. We assessed the spatial and temporal distribution of individuals (males, females, juveniles) and explored potential relationships with environmental variables. Field activities encompassed bimonthly surveys in three semi-permanent ponds, each one during approximately two days and two nights, from August 2004 to July 2005. We recorded differences in the sites used by males, females and juveniles, with males occupying deeper and more distant places from the border. The temporal distributions of individuals, calling sites and amplectant pairs indicated that the reproductive activity of P. minuta is related to some of the studied abiotic factors. Calling males presented statistical differences in relation to non-calling males for all daily abiotic variables analyzed (air temperature, water temperature, relative humidity and rainfall), as well as to monthly temperature and rainfall. The number of active males, females and juveniles was influenced by at least one of the daily or monthly environmental variables analyzed. We conclude that the reproduction in this species is seasonal and may be partially determined by abiotic factors.

New records, distribution and status of six seabird species in Brazil

Distribution records of poorly-known species are currently the most explored theme in the Brazilian seabird literature. If properly evaluated, this kind of information can improve our knowledge on distribution, migration and status of occurrence of these species. In this note we present new records for six species of poorly-known seabirds in the Brazilian coast, reviewing distribution records and defining their status of occurrence in the country. We consider Chionis albus (Gmelin, 1789) a pseudo-vagrant in Brazil and define its status as a scarce seasonal visitor from southern South America. We present the first records of Leucophaeus atricilla (Linnaeus, 1758) for Trindade Island, and of Leucophaeus pipixcan (Wagler, 1831) for the state of Rio Grande do Sul, and determined that the former is a vagrant in eastern Brazil and the latter a vagrant across the country. Anous stolidus (Linnaeus, 1758) is a vagrant in southernmost Brazil. We were unable to determine if records of Chlidonias niger (Linnaeus, 1758) for Brazil and southern South America refer to vagrancy or pseudo-vagrancy. Additionally, we verified the occurrence of breeding individuals of Anous minutus Boie, 1844 on Martin Vaz Island and confirmed that there is no evidence of breeding on neighboring Trindade Island.

Clutch size in the small-sized lizard Eurolophosaurus nanuzae (Tropiduridae): does it vary along the geographic distribution of the species?

We studied life history traits of females of the lizard Eurolophosaurus nanuzae (Rodrigues, 1981), an endemic species of rock outcrop habitats in southeastern Brazil. During October 2002 and 2003 we sampled three populations in sites that encompass the meridional portion of the geographic range of the species. Clutch size varied from one to three eggs, with most females carrying two eggs. Clutch size did not vary among populations, but was correlated to female body size. Only larger females produced clutches of three eggs. Females of the small-sized E. nanuzae produce eggs as large as those of medium-sized tropidurids, thus investing a considerable amount of energy to produce clutches resulting in high values of relative clutch mass.